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Target_Summary_ID stringclasses 200 values	Target_Sentence_Index stringlengths 4 7	External stringclasses 2 values	Target_Sentence stringlengths 19 592	Original_Abstract stringclasses 200 values
t136	t136_4	yes	An incorrect tuberculosis diagnosis may result in anxiety and unnecessary treatment.	The lateral flow urine lipoarabinomannan (LF‐LAM) assay Alere Determine™ TB LAM Ag is recommended by the World Health Organization (WHO) to help detect active tuberculosis in HIV‐positive people with severe HIV disease. This review update asks the question, "does new evidence justify the use of LF‐LAM in a broader group of people?”, and is part of the WHO process for updating guidance on the use of LF‐LAM. Objectives To assess the accuracy of LF‐LAM for the diagnosis of active tuberculosis among HIV‐positive adults with signs and symptoms of tuberculosis (symptomatic participants) and among HIV‐positive adults irrespective of signs and symptoms of tuberculosis (unselected participants not assessed for tuberculosis signs and symptoms). The proposed role for LF‐LAM is as an add on to clinical judgement and with other tests to assist in diagnosing tuberculosis. Search methods We searched the Cochrane Infectious Diseases Group Specialized Register; MEDLINE, Embase, Science Citation Index, Web of Science, Latin American Caribbean Health Sciences Literature, Scopus, the WHO International Clinical Trials Registry Platform, the International Standard Randomized Controlled Trial Number Registry, and ProQuest, without language restriction to 11 May 2018. Selection criteria Randomized trials, cross‐sectional, and observational cohort studies that evaluated LF‐LAM for active tuberculosis (pulmonary and extrapulmonary) in HIV‐positive adults. We included studies that used the manufacturer's recommended threshold for test positivity, either the updated reference card with four bands (grade 1 of 4) or the corresponding prior reference card grade with five bands (grade 2 of 5). The reference standard was culture or nucleic acid amplification test from any body site (microbiological). We considered a higher quality reference standard to be one in which two or more specimen types were evaluated for tuberculosis diagnosis and a lower quality reference standard to be one in which only one specimen type was evaluated. Data collection and analysis Two review authors independently extracted data using a standardized form and REDCap electronic data capture tools. We appraised the quality of studies using the Quality Assessment of Diagnostic Accuracy Studies‐2 (QUADAS‐2) tool and performed meta‐analyses to estimate pooled sensitivity and specificity using a bivariate random‐effects model and a Bayesian approach. We analyzed studies enrolling strictly symptomatic participants separately from those enrolling unselected participants. We investigated pre‐defined sources of heterogeneity including the influence of CD4 count and clinical setting on the accuracy estimates. We assessed the certainty of the evidence using the GRADE approach. We included 15 unique studies (nine new studies and six studies from the original review that met the inclusion criteria): eight studies among symptomatic adults and seven studies among unselected adults. All studies were conducted in low‐ or middle‐income countries. Risk of bias was high in the patient selection and reference standard domains, mainly because studies excluded participants unable to produce sputum and used a lower quality reference standard. Participants with tuberculosis symptoms LF‐LAM pooled sensitivity (95% credible interval (CrI) ) was 42% (31% to 55%) (moderate‐certainty evidence) and pooled specificity was 91% (85% to 95%) (very low‐certainty evidence), (8 studies, 3449 participants, 37% with tuberculosis). For a population of 1000 people where 300 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 189 to be LF‐LAM positive: of these, 63 (33%) would not have tuberculosis (false‐positives); and 811 to be LF‐LAM negative: of these, 174 (21%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 52% (40% to 64%) among inpatients versus 29% (17% to 47%) among outpatients; and pooled specificity was 87% (78% to 93%) among inpatients versus 96% (91% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. Unselected participants not assessed for signs and symptoms of tuberculosis LF‐LAM pooled sensitivity was 35% (22% to 50%), (moderate‐certainty evidence) and pooled specificity was 95% (89% to 96%), (low‐certainty evidence), (7 studies, 3365 participants, 13% with tuberculosis). For a population of 1000 people where 100 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 80 to be LF‐LAM positive: of these, 45 (56%) would not have tuberculosis (false‐positives); and 920 to be LF‐LAM negative: of these, 65 (7%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 62% (41% to 83%) among inpatients versus 31% (18% to 47%) among outpatients; pooled specificity was 84% (48% to 96%) among inpatients versus 95% (87% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. We found that LF‐LAM has a sensitivity of 42% to diagnose tuberculosis in HIV‐positive individuals with tuberculosis symptoms and 35% in HIV‐positive individuals not assessed for tuberculosis symptoms, consistent with findings reported previously. Regardless of how people are enrolled, sensitivity is higher in inpatients and those with lower CD4 cell, but a concomitant lower specificity. As a simple point‐of‐care test that does not depend upon sputum evaluation, LF‐LAM may assist with the diagnosis of tuberculosis, particularly when a sputum specimen cannot be produced. 17 October 2019 Up to date All studies incorporated from most recent search All studies identified during the most recent search (11 May, 2018) have been incorporated in the review, and no ongoing studies identified.
t136	t136_5	no	To find out how accurate LF‐LAM is for diagnosing tuberculosis in HIV‐positive people with tuberculosis symptoms (symptomatic participants) and those not assessed for tuberculosis symptoms (unselected participants).	The lateral flow urine lipoarabinomannan (LF‐LAM) assay Alere Determine™ TB LAM Ag is recommended by the World Health Organization (WHO) to help detect active tuberculosis in HIV‐positive people with severe HIV disease. This review update asks the question, "does new evidence justify the use of LF‐LAM in a broader group of people?”, and is part of the WHO process for updating guidance on the use of LF‐LAM. Objectives To assess the accuracy of LF‐LAM for the diagnosis of active tuberculosis among HIV‐positive adults with signs and symptoms of tuberculosis (symptomatic participants) and among HIV‐positive adults irrespective of signs and symptoms of tuberculosis (unselected participants not assessed for tuberculosis signs and symptoms). The proposed role for LF‐LAM is as an add on to clinical judgement and with other tests to assist in diagnosing tuberculosis. Search methods We searched the Cochrane Infectious Diseases Group Specialized Register; MEDLINE, Embase, Science Citation Index, Web of Science, Latin American Caribbean Health Sciences Literature, Scopus, the WHO International Clinical Trials Registry Platform, the International Standard Randomized Controlled Trial Number Registry, and ProQuest, without language restriction to 11 May 2018. Selection criteria Randomized trials, cross‐sectional, and observational cohort studies that evaluated LF‐LAM for active tuberculosis (pulmonary and extrapulmonary) in HIV‐positive adults. We included studies that used the manufacturer's recommended threshold for test positivity, either the updated reference card with four bands (grade 1 of 4) or the corresponding prior reference card grade with five bands (grade 2 of 5). The reference standard was culture or nucleic acid amplification test from any body site (microbiological). We considered a higher quality reference standard to be one in which two or more specimen types were evaluated for tuberculosis diagnosis and a lower quality reference standard to be one in which only one specimen type was evaluated. Data collection and analysis Two review authors independently extracted data using a standardized form and REDCap electronic data capture tools. We appraised the quality of studies using the Quality Assessment of Diagnostic Accuracy Studies‐2 (QUADAS‐2) tool and performed meta‐analyses to estimate pooled sensitivity and specificity using a bivariate random‐effects model and a Bayesian approach. We analyzed studies enrolling strictly symptomatic participants separately from those enrolling unselected participants. We investigated pre‐defined sources of heterogeneity including the influence of CD4 count and clinical setting on the accuracy estimates. We assessed the certainty of the evidence using the GRADE approach. We included 15 unique studies (nine new studies and six studies from the original review that met the inclusion criteria): eight studies among symptomatic adults and seven studies among unselected adults. All studies were conducted in low‐ or middle‐income countries. Risk of bias was high in the patient selection and reference standard domains, mainly because studies excluded participants unable to produce sputum and used a lower quality reference standard. Participants with tuberculosis symptoms LF‐LAM pooled sensitivity (95% credible interval (CrI) ) was 42% (31% to 55%) (moderate‐certainty evidence) and pooled specificity was 91% (85% to 95%) (very low‐certainty evidence), (8 studies, 3449 participants, 37% with tuberculosis). For a population of 1000 people where 300 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 189 to be LF‐LAM positive: of these, 63 (33%) would not have tuberculosis (false‐positives); and 811 to be LF‐LAM negative: of these, 174 (21%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 52% (40% to 64%) among inpatients versus 29% (17% to 47%) among outpatients; and pooled specificity was 87% (78% to 93%) among inpatients versus 96% (91% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. Unselected participants not assessed for signs and symptoms of tuberculosis LF‐LAM pooled sensitivity was 35% (22% to 50%), (moderate‐certainty evidence) and pooled specificity was 95% (89% to 96%), (low‐certainty evidence), (7 studies, 3365 participants, 13% with tuberculosis). For a population of 1000 people where 100 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 80 to be LF‐LAM positive: of these, 45 (56%) would not have tuberculosis (false‐positives); and 920 to be LF‐LAM negative: of these, 65 (7%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 62% (41% to 83%) among inpatients versus 31% (18% to 47%) among outpatients; pooled specificity was 84% (48% to 96%) among inpatients versus 95% (87% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. We found that LF‐LAM has a sensitivity of 42% to diagnose tuberculosis in HIV‐positive individuals with tuberculosis symptoms and 35% in HIV‐positive individuals not assessed for tuberculosis symptoms, consistent with findings reported previously. Regardless of how people are enrolled, sensitivity is higher in inpatients and those with lower CD4 cell, but a concomitant lower specificity. As a simple point‐of‐care test that does not depend upon sputum evaluation, LF‐LAM may assist with the diagnosis of tuberculosis, particularly when a sputum specimen cannot be produced. 17 October 2019 Up to date All studies incorporated from most recent search All studies identified during the most recent search (11 May, 2018) have been incorporated in the review, and no ongoing studies identified.
t136	t136_6	yes	LF‐LAM is a commercially available point‐of‐care test that detects lipoarabinomannan (LAM), a component of the bacterial cell walls, present in some people with active tuberculosis.	The lateral flow urine lipoarabinomannan (LF‐LAM) assay Alere Determine™ TB LAM Ag is recommended by the World Health Organization (WHO) to help detect active tuberculosis in HIV‐positive people with severe HIV disease. This review update asks the question, "does new evidence justify the use of LF‐LAM in a broader group of people?”, and is part of the WHO process for updating guidance on the use of LF‐LAM. Objectives To assess the accuracy of LF‐LAM for the diagnosis of active tuberculosis among HIV‐positive adults with signs and symptoms of tuberculosis (symptomatic participants) and among HIV‐positive adults irrespective of signs and symptoms of tuberculosis (unselected participants not assessed for tuberculosis signs and symptoms). The proposed role for LF‐LAM is as an add on to clinical judgement and with other tests to assist in diagnosing tuberculosis. Search methods We searched the Cochrane Infectious Diseases Group Specialized Register; MEDLINE, Embase, Science Citation Index, Web of Science, Latin American Caribbean Health Sciences Literature, Scopus, the WHO International Clinical Trials Registry Platform, the International Standard Randomized Controlled Trial Number Registry, and ProQuest, without language restriction to 11 May 2018. Selection criteria Randomized trials, cross‐sectional, and observational cohort studies that evaluated LF‐LAM for active tuberculosis (pulmonary and extrapulmonary) in HIV‐positive adults. We included studies that used the manufacturer's recommended threshold for test positivity, either the updated reference card with four bands (grade 1 of 4) or the corresponding prior reference card grade with five bands (grade 2 of 5). The reference standard was culture or nucleic acid amplification test from any body site (microbiological). We considered a higher quality reference standard to be one in which two or more specimen types were evaluated for tuberculosis diagnosis and a lower quality reference standard to be one in which only one specimen type was evaluated. Data collection and analysis Two review authors independently extracted data using a standardized form and REDCap electronic data capture tools. We appraised the quality of studies using the Quality Assessment of Diagnostic Accuracy Studies‐2 (QUADAS‐2) tool and performed meta‐analyses to estimate pooled sensitivity and specificity using a bivariate random‐effects model and a Bayesian approach. We analyzed studies enrolling strictly symptomatic participants separately from those enrolling unselected participants. We investigated pre‐defined sources of heterogeneity including the influence of CD4 count and clinical setting on the accuracy estimates. We assessed the certainty of the evidence using the GRADE approach. We included 15 unique studies (nine new studies and six studies from the original review that met the inclusion criteria): eight studies among symptomatic adults and seven studies among unselected adults. All studies were conducted in low‐ or middle‐income countries. Risk of bias was high in the patient selection and reference standard domains, mainly because studies excluded participants unable to produce sputum and used a lower quality reference standard. Participants with tuberculosis symptoms LF‐LAM pooled sensitivity (95% credible interval (CrI) ) was 42% (31% to 55%) (moderate‐certainty evidence) and pooled specificity was 91% (85% to 95%) (very low‐certainty evidence), (8 studies, 3449 participants, 37% with tuberculosis). For a population of 1000 people where 300 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 189 to be LF‐LAM positive: of these, 63 (33%) would not have tuberculosis (false‐positives); and 811 to be LF‐LAM negative: of these, 174 (21%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 52% (40% to 64%) among inpatients versus 29% (17% to 47%) among outpatients; and pooled specificity was 87% (78% to 93%) among inpatients versus 96% (91% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. Unselected participants not assessed for signs and symptoms of tuberculosis LF‐LAM pooled sensitivity was 35% (22% to 50%), (moderate‐certainty evidence) and pooled specificity was 95% (89% to 96%), (low‐certainty evidence), (7 studies, 3365 participants, 13% with tuberculosis). For a population of 1000 people where 100 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 80 to be LF‐LAM positive: of these, 45 (56%) would not have tuberculosis (false‐positives); and 920 to be LF‐LAM negative: of these, 65 (7%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 62% (41% to 83%) among inpatients versus 31% (18% to 47%) among outpatients; pooled specificity was 84% (48% to 96%) among inpatients versus 95% (87% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. We found that LF‐LAM has a sensitivity of 42% to diagnose tuberculosis in HIV‐positive individuals with tuberculosis symptoms and 35% in HIV‐positive individuals not assessed for tuberculosis symptoms, consistent with findings reported previously. Regardless of how people are enrolled, sensitivity is higher in inpatients and those with lower CD4 cell, but a concomitant lower specificity. As a simple point‐of‐care test that does not depend upon sputum evaluation, LF‐LAM may assist with the diagnosis of tuberculosis, particularly when a sputum specimen cannot be produced. 17 October 2019 Up to date All studies incorporated from most recent search All studies identified during the most recent search (11 May, 2018) have been incorporated in the review, and no ongoing studies identified.
t136	t136_7	no	LF‐LAM results were measured against culture or molecular tests (benchmark).	The lateral flow urine lipoarabinomannan (LF‐LAM) assay Alere Determine™ TB LAM Ag is recommended by the World Health Organization (WHO) to help detect active tuberculosis in HIV‐positive people with severe HIV disease. This review update asks the question, "does new evidence justify the use of LF‐LAM in a broader group of people?”, and is part of the WHO process for updating guidance on the use of LF‐LAM. Objectives To assess the accuracy of LF‐LAM for the diagnosis of active tuberculosis among HIV‐positive adults with signs and symptoms of tuberculosis (symptomatic participants) and among HIV‐positive adults irrespective of signs and symptoms of tuberculosis (unselected participants not assessed for tuberculosis signs and symptoms). The proposed role for LF‐LAM is as an add on to clinical judgement and with other tests to assist in diagnosing tuberculosis. Search methods We searched the Cochrane Infectious Diseases Group Specialized Register; MEDLINE, Embase, Science Citation Index, Web of Science, Latin American Caribbean Health Sciences Literature, Scopus, the WHO International Clinical Trials Registry Platform, the International Standard Randomized Controlled Trial Number Registry, and ProQuest, without language restriction to 11 May 2018. Selection criteria Randomized trials, cross‐sectional, and observational cohort studies that evaluated LF‐LAM for active tuberculosis (pulmonary and extrapulmonary) in HIV‐positive adults. We included studies that used the manufacturer's recommended threshold for test positivity, either the updated reference card with four bands (grade 1 of 4) or the corresponding prior reference card grade with five bands (grade 2 of 5). The reference standard was culture or nucleic acid amplification test from any body site (microbiological). We considered a higher quality reference standard to be one in which two or more specimen types were evaluated for tuberculosis diagnosis and a lower quality reference standard to be one in which only one specimen type was evaluated. Data collection and analysis Two review authors independently extracted data using a standardized form and REDCap electronic data capture tools. We appraised the quality of studies using the Quality Assessment of Diagnostic Accuracy Studies‐2 (QUADAS‐2) tool and performed meta‐analyses to estimate pooled sensitivity and specificity using a bivariate random‐effects model and a Bayesian approach. We analyzed studies enrolling strictly symptomatic participants separately from those enrolling unselected participants. We investigated pre‐defined sources of heterogeneity including the influence of CD4 count and clinical setting on the accuracy estimates. We assessed the certainty of the evidence using the GRADE approach. We included 15 unique studies (nine new studies and six studies from the original review that met the inclusion criteria): eight studies among symptomatic adults and seven studies among unselected adults. All studies were conducted in low‐ or middle‐income countries. Risk of bias was high in the patient selection and reference standard domains, mainly because studies excluded participants unable to produce sputum and used a lower quality reference standard. Participants with tuberculosis symptoms LF‐LAM pooled sensitivity (95% credible interval (CrI) ) was 42% (31% to 55%) (moderate‐certainty evidence) and pooled specificity was 91% (85% to 95%) (very low‐certainty evidence), (8 studies, 3449 participants, 37% with tuberculosis). For a population of 1000 people where 300 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 189 to be LF‐LAM positive: of these, 63 (33%) would not have tuberculosis (false‐positives); and 811 to be LF‐LAM negative: of these, 174 (21%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 52% (40% to 64%) among inpatients versus 29% (17% to 47%) among outpatients; and pooled specificity was 87% (78% to 93%) among inpatients versus 96% (91% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. Unselected participants not assessed for signs and symptoms of tuberculosis LF‐LAM pooled sensitivity was 35% (22% to 50%), (moderate‐certainty evidence) and pooled specificity was 95% (89% to 96%), (low‐certainty evidence), (7 studies, 3365 participants, 13% with tuberculosis). For a population of 1000 people where 100 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 80 to be LF‐LAM positive: of these, 45 (56%) would not have tuberculosis (false‐positives); and 920 to be LF‐LAM negative: of these, 65 (7%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 62% (41% to 83%) among inpatients versus 31% (18% to 47%) among outpatients; pooled specificity was 84% (48% to 96%) among inpatients versus 95% (87% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. We found that LF‐LAM has a sensitivity of 42% to diagnose tuberculosis in HIV‐positive individuals with tuberculosis symptoms and 35% in HIV‐positive individuals not assessed for tuberculosis symptoms, consistent with findings reported previously. Regardless of how people are enrolled, sensitivity is higher in inpatients and those with lower CD4 cell, but a concomitant lower specificity. As a simple point‐of‐care test that does not depend upon sputum evaluation, LF‐LAM may assist with the diagnosis of tuberculosis, particularly when a sputum specimen cannot be produced. 17 October 2019 Up to date All studies incorporated from most recent search All studies identified during the most recent search (11 May, 2018) have been incorporated in the review, and no ongoing studies identified.
t136	t136_8	no	Fifteen studies: eight studies evaluated LF‐LAM for tuberculosis among symptomatic participants and seven studies among unselected participants.	The lateral flow urine lipoarabinomannan (LF‐LAM) assay Alere Determine™ TB LAM Ag is recommended by the World Health Organization (WHO) to help detect active tuberculosis in HIV‐positive people with severe HIV disease. This review update asks the question, "does new evidence justify the use of LF‐LAM in a broader group of people?”, and is part of the WHO process for updating guidance on the use of LF‐LAM. Objectives To assess the accuracy of LF‐LAM for the diagnosis of active tuberculosis among HIV‐positive adults with signs and symptoms of tuberculosis (symptomatic participants) and among HIV‐positive adults irrespective of signs and symptoms of tuberculosis (unselected participants not assessed for tuberculosis signs and symptoms). The proposed role for LF‐LAM is as an add on to clinical judgement and with other tests to assist in diagnosing tuberculosis. Search methods We searched the Cochrane Infectious Diseases Group Specialized Register; MEDLINE, Embase, Science Citation Index, Web of Science, Latin American Caribbean Health Sciences Literature, Scopus, the WHO International Clinical Trials Registry Platform, the International Standard Randomized Controlled Trial Number Registry, and ProQuest, without language restriction to 11 May 2018. Selection criteria Randomized trials, cross‐sectional, and observational cohort studies that evaluated LF‐LAM for active tuberculosis (pulmonary and extrapulmonary) in HIV‐positive adults. We included studies that used the manufacturer's recommended threshold for test positivity, either the updated reference card with four bands (grade 1 of 4) or the corresponding prior reference card grade with five bands (grade 2 of 5). The reference standard was culture or nucleic acid amplification test from any body site (microbiological). We considered a higher quality reference standard to be one in which two or more specimen types were evaluated for tuberculosis diagnosis and a lower quality reference standard to be one in which only one specimen type was evaluated. Data collection and analysis Two review authors independently extracted data using a standardized form and REDCap electronic data capture tools. We appraised the quality of studies using the Quality Assessment of Diagnostic Accuracy Studies‐2 (QUADAS‐2) tool and performed meta‐analyses to estimate pooled sensitivity and specificity using a bivariate random‐effects model and a Bayesian approach. We analyzed studies enrolling strictly symptomatic participants separately from those enrolling unselected participants. We investigated pre‐defined sources of heterogeneity including the influence of CD4 count and clinical setting on the accuracy estimates. We assessed the certainty of the evidence using the GRADE approach. We included 15 unique studies (nine new studies and six studies from the original review that met the inclusion criteria): eight studies among symptomatic adults and seven studies among unselected adults. All studies were conducted in low‐ or middle‐income countries. Risk of bias was high in the patient selection and reference standard domains, mainly because studies excluded participants unable to produce sputum and used a lower quality reference standard. Participants with tuberculosis symptoms LF‐LAM pooled sensitivity (95% credible interval (CrI) ) was 42% (31% to 55%) (moderate‐certainty evidence) and pooled specificity was 91% (85% to 95%) (very low‐certainty evidence), (8 studies, 3449 participants, 37% with tuberculosis). For a population of 1000 people where 300 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 189 to be LF‐LAM positive: of these, 63 (33%) would not have tuberculosis (false‐positives); and 811 to be LF‐LAM negative: of these, 174 (21%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 52% (40% to 64%) among inpatients versus 29% (17% to 47%) among outpatients; and pooled specificity was 87% (78% to 93%) among inpatients versus 96% (91% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. Unselected participants not assessed for signs and symptoms of tuberculosis LF‐LAM pooled sensitivity was 35% (22% to 50%), (moderate‐certainty evidence) and pooled specificity was 95% (89% to 96%), (low‐certainty evidence), (7 studies, 3365 participants, 13% with tuberculosis). For a population of 1000 people where 100 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 80 to be LF‐LAM positive: of these, 45 (56%) would not have tuberculosis (false‐positives); and 920 to be LF‐LAM negative: of these, 65 (7%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 62% (41% to 83%) among inpatients versus 31% (18% to 47%) among outpatients; pooled specificity was 84% (48% to 96%) among inpatients versus 95% (87% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. We found that LF‐LAM has a sensitivity of 42% to diagnose tuberculosis in HIV‐positive individuals with tuberculosis symptoms and 35% in HIV‐positive individuals not assessed for tuberculosis symptoms, consistent with findings reported previously. Regardless of how people are enrolled, sensitivity is higher in inpatients and those with lower CD4 cell, but a concomitant lower specificity. As a simple point‐of‐care test that does not depend upon sputum evaluation, LF‐LAM may assist with the diagnosis of tuberculosis, particularly when a sputum specimen cannot be produced. 17 October 2019 Up to date All studies incorporated from most recent search All studies identified during the most recent search (11 May, 2018) have been incorporated in the review, and no ongoing studies identified.
t136	t136_9	no	Tuberculosis diagnosis among symptomatic participants: LF‐LAM registered positive in 42% (sensitivity) of people who actually had tuberculosis and did not register positive in 91% of people who were actually negative (specificity).	The lateral flow urine lipoarabinomannan (LF‐LAM) assay Alere Determine™ TB LAM Ag is recommended by the World Health Organization (WHO) to help detect active tuberculosis in HIV‐positive people with severe HIV disease. This review update asks the question, "does new evidence justify the use of LF‐LAM in a broader group of people?”, and is part of the WHO process for updating guidance on the use of LF‐LAM. Objectives To assess the accuracy of LF‐LAM for the diagnosis of active tuberculosis among HIV‐positive adults with signs and symptoms of tuberculosis (symptomatic participants) and among HIV‐positive adults irrespective of signs and symptoms of tuberculosis (unselected participants not assessed for tuberculosis signs and symptoms). The proposed role for LF‐LAM is as an add on to clinical judgement and with other tests to assist in diagnosing tuberculosis. Search methods We searched the Cochrane Infectious Diseases Group Specialized Register; MEDLINE, Embase, Science Citation Index, Web of Science, Latin American Caribbean Health Sciences Literature, Scopus, the WHO International Clinical Trials Registry Platform, the International Standard Randomized Controlled Trial Number Registry, and ProQuest, without language restriction to 11 May 2018. Selection criteria Randomized trials, cross‐sectional, and observational cohort studies that evaluated LF‐LAM for active tuberculosis (pulmonary and extrapulmonary) in HIV‐positive adults. We included studies that used the manufacturer's recommended threshold for test positivity, either the updated reference card with four bands (grade 1 of 4) or the corresponding prior reference card grade with five bands (grade 2 of 5). The reference standard was culture or nucleic acid amplification test from any body site (microbiological). We considered a higher quality reference standard to be one in which two or more specimen types were evaluated for tuberculosis diagnosis and a lower quality reference standard to be one in which only one specimen type was evaluated. Data collection and analysis Two review authors independently extracted data using a standardized form and REDCap electronic data capture tools. We appraised the quality of studies using the Quality Assessment of Diagnostic Accuracy Studies‐2 (QUADAS‐2) tool and performed meta‐analyses to estimate pooled sensitivity and specificity using a bivariate random‐effects model and a Bayesian approach. We analyzed studies enrolling strictly symptomatic participants separately from those enrolling unselected participants. We investigated pre‐defined sources of heterogeneity including the influence of CD4 count and clinical setting on the accuracy estimates. We assessed the certainty of the evidence using the GRADE approach. We included 15 unique studies (nine new studies and six studies from the original review that met the inclusion criteria): eight studies among symptomatic adults and seven studies among unselected adults. All studies were conducted in low‐ or middle‐income countries. Risk of bias was high in the patient selection and reference standard domains, mainly because studies excluded participants unable to produce sputum and used a lower quality reference standard. Participants with tuberculosis symptoms LF‐LAM pooled sensitivity (95% credible interval (CrI) ) was 42% (31% to 55%) (moderate‐certainty evidence) and pooled specificity was 91% (85% to 95%) (very low‐certainty evidence), (8 studies, 3449 participants, 37% with tuberculosis). For a population of 1000 people where 300 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 189 to be LF‐LAM positive: of these, 63 (33%) would not have tuberculosis (false‐positives); and 811 to be LF‐LAM negative: of these, 174 (21%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 52% (40% to 64%) among inpatients versus 29% (17% to 47%) among outpatients; and pooled specificity was 87% (78% to 93%) among inpatients versus 96% (91% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. Unselected participants not assessed for signs and symptoms of tuberculosis LF‐LAM pooled sensitivity was 35% (22% to 50%), (moderate‐certainty evidence) and pooled specificity was 95% (89% to 96%), (low‐certainty evidence), (7 studies, 3365 participants, 13% with tuberculosis). For a population of 1000 people where 100 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 80 to be LF‐LAM positive: of these, 45 (56%) would not have tuberculosis (false‐positives); and 920 to be LF‐LAM negative: of these, 65 (7%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 62% (41% to 83%) among inpatients versus 31% (18% to 47%) among outpatients; pooled specificity was 84% (48% to 96%) among inpatients versus 95% (87% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. We found that LF‐LAM has a sensitivity of 42% to diagnose tuberculosis in HIV‐positive individuals with tuberculosis symptoms and 35% in HIV‐positive individuals not assessed for tuberculosis symptoms, consistent with findings reported previously. Regardless of how people are enrolled, sensitivity is higher in inpatients and those with lower CD4 cell, but a concomitant lower specificity. As a simple point‐of‐care test that does not depend upon sputum evaluation, LF‐LAM may assist with the diagnosis of tuberculosis, particularly when a sputum specimen cannot be produced. 17 October 2019 Up to date All studies incorporated from most recent search All studies identified during the most recent search (11 May, 2018) have been incorporated in the review, and no ongoing studies identified.
t136	t136_10	no	Tuberculosis diagnosis among unselected participants: LF‐LAM sensitivity was 35% and specificity 95%.	The lateral flow urine lipoarabinomannan (LF‐LAM) assay Alere Determine™ TB LAM Ag is recommended by the World Health Organization (WHO) to help detect active tuberculosis in HIV‐positive people with severe HIV disease. This review update asks the question, "does new evidence justify the use of LF‐LAM in a broader group of people?”, and is part of the WHO process for updating guidance on the use of LF‐LAM. Objectives To assess the accuracy of LF‐LAM for the diagnosis of active tuberculosis among HIV‐positive adults with signs and symptoms of tuberculosis (symptomatic participants) and among HIV‐positive adults irrespective of signs and symptoms of tuberculosis (unselected participants not assessed for tuberculosis signs and symptoms). The proposed role for LF‐LAM is as an add on to clinical judgement and with other tests to assist in diagnosing tuberculosis. Search methods We searched the Cochrane Infectious Diseases Group Specialized Register; MEDLINE, Embase, Science Citation Index, Web of Science, Latin American Caribbean Health Sciences Literature, Scopus, the WHO International Clinical Trials Registry Platform, the International Standard Randomized Controlled Trial Number Registry, and ProQuest, without language restriction to 11 May 2018. Selection criteria Randomized trials, cross‐sectional, and observational cohort studies that evaluated LF‐LAM for active tuberculosis (pulmonary and extrapulmonary) in HIV‐positive adults. We included studies that used the manufacturer's recommended threshold for test positivity, either the updated reference card with four bands (grade 1 of 4) or the corresponding prior reference card grade with five bands (grade 2 of 5). The reference standard was culture or nucleic acid amplification test from any body site (microbiological). We considered a higher quality reference standard to be one in which two or more specimen types were evaluated for tuberculosis diagnosis and a lower quality reference standard to be one in which only one specimen type was evaluated. Data collection and analysis Two review authors independently extracted data using a standardized form and REDCap electronic data capture tools. We appraised the quality of studies using the Quality Assessment of Diagnostic Accuracy Studies‐2 (QUADAS‐2) tool and performed meta‐analyses to estimate pooled sensitivity and specificity using a bivariate random‐effects model and a Bayesian approach. We analyzed studies enrolling strictly symptomatic participants separately from those enrolling unselected participants. We investigated pre‐defined sources of heterogeneity including the influence of CD4 count and clinical setting on the accuracy estimates. We assessed the certainty of the evidence using the GRADE approach. We included 15 unique studies (nine new studies and six studies from the original review that met the inclusion criteria): eight studies among symptomatic adults and seven studies among unselected adults. All studies were conducted in low‐ or middle‐income countries. Risk of bias was high in the patient selection and reference standard domains, mainly because studies excluded participants unable to produce sputum and used a lower quality reference standard. Participants with tuberculosis symptoms LF‐LAM pooled sensitivity (95% credible interval (CrI) ) was 42% (31% to 55%) (moderate‐certainty evidence) and pooled specificity was 91% (85% to 95%) (very low‐certainty evidence), (8 studies, 3449 participants, 37% with tuberculosis). For a population of 1000 people where 300 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 189 to be LF‐LAM positive: of these, 63 (33%) would not have tuberculosis (false‐positives); and 811 to be LF‐LAM negative: of these, 174 (21%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 52% (40% to 64%) among inpatients versus 29% (17% to 47%) among outpatients; and pooled specificity was 87% (78% to 93%) among inpatients versus 96% (91% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. Unselected participants not assessed for signs and symptoms of tuberculosis LF‐LAM pooled sensitivity was 35% (22% to 50%), (moderate‐certainty evidence) and pooled specificity was 95% (89% to 96%), (low‐certainty evidence), (7 studies, 3365 participants, 13% with tuberculosis). For a population of 1000 people where 100 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 80 to be LF‐LAM positive: of these, 45 (56%) would not have tuberculosis (false‐positives); and 920 to be LF‐LAM negative: of these, 65 (7%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 62% (41% to 83%) among inpatients versus 31% (18% to 47%) among outpatients; pooled specificity was 84% (48% to 96%) among inpatients versus 95% (87% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. We found that LF‐LAM has a sensitivity of 42% to diagnose tuberculosis in HIV‐positive individuals with tuberculosis symptoms and 35% in HIV‐positive individuals not assessed for tuberculosis symptoms, consistent with findings reported previously. Regardless of how people are enrolled, sensitivity is higher in inpatients and those with lower CD4 cell, but a concomitant lower specificity. As a simple point‐of‐care test that does not depend upon sputum evaluation, LF‐LAM may assist with the diagnosis of tuberculosis, particularly when a sputum specimen cannot be produced. 17 October 2019 Up to date All studies incorporated from most recent search All studies identified during the most recent search (11 May, 2018) have been incorporated in the review, and no ongoing studies identified.
t136	t136_11	no	Several studies excluded participants who could not produce sputum and most studies relied on a lower quality benchmark.	The lateral flow urine lipoarabinomannan (LF‐LAM) assay Alere Determine™ TB LAM Ag is recommended by the World Health Organization (WHO) to help detect active tuberculosis in HIV‐positive people with severe HIV disease. This review update asks the question, "does new evidence justify the use of LF‐LAM in a broader group of people?”, and is part of the WHO process for updating guidance on the use of LF‐LAM. Objectives To assess the accuracy of LF‐LAM for the diagnosis of active tuberculosis among HIV‐positive adults with signs and symptoms of tuberculosis (symptomatic participants) and among HIV‐positive adults irrespective of signs and symptoms of tuberculosis (unselected participants not assessed for tuberculosis signs and symptoms). The proposed role for LF‐LAM is as an add on to clinical judgement and with other tests to assist in diagnosing tuberculosis. Search methods We searched the Cochrane Infectious Diseases Group Specialized Register; MEDLINE, Embase, Science Citation Index, Web of Science, Latin American Caribbean Health Sciences Literature, Scopus, the WHO International Clinical Trials Registry Platform, the International Standard Randomized Controlled Trial Number Registry, and ProQuest, without language restriction to 11 May 2018. Selection criteria Randomized trials, cross‐sectional, and observational cohort studies that evaluated LF‐LAM for active tuberculosis (pulmonary and extrapulmonary) in HIV‐positive adults. We included studies that used the manufacturer's recommended threshold for test positivity, either the updated reference card with four bands (grade 1 of 4) or the corresponding prior reference card grade with five bands (grade 2 of 5). The reference standard was culture or nucleic acid amplification test from any body site (microbiological). We considered a higher quality reference standard to be one in which two or more specimen types were evaluated for tuberculosis diagnosis and a lower quality reference standard to be one in which only one specimen type was evaluated. Data collection and analysis Two review authors independently extracted data using a standardized form and REDCap electronic data capture tools. We appraised the quality of studies using the Quality Assessment of Diagnostic Accuracy Studies‐2 (QUADAS‐2) tool and performed meta‐analyses to estimate pooled sensitivity and specificity using a bivariate random‐effects model and a Bayesian approach. We analyzed studies enrolling strictly symptomatic participants separately from those enrolling unselected participants. We investigated pre‐defined sources of heterogeneity including the influence of CD4 count and clinical setting on the accuracy estimates. We assessed the certainty of the evidence using the GRADE approach. We included 15 unique studies (nine new studies and six studies from the original review that met the inclusion criteria): eight studies among symptomatic adults and seven studies among unselected adults. All studies were conducted in low‐ or middle‐income countries. Risk of bias was high in the patient selection and reference standard domains, mainly because studies excluded participants unable to produce sputum and used a lower quality reference standard. Participants with tuberculosis symptoms LF‐LAM pooled sensitivity (95% credible interval (CrI) ) was 42% (31% to 55%) (moderate‐certainty evidence) and pooled specificity was 91% (85% to 95%) (very low‐certainty evidence), (8 studies, 3449 participants, 37% with tuberculosis). For a population of 1000 people where 300 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 189 to be LF‐LAM positive: of these, 63 (33%) would not have tuberculosis (false‐positives); and 811 to be LF‐LAM negative: of these, 174 (21%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 52% (40% to 64%) among inpatients versus 29% (17% to 47%) among outpatients; and pooled specificity was 87% (78% to 93%) among inpatients versus 96% (91% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. Unselected participants not assessed for signs and symptoms of tuberculosis LF‐LAM pooled sensitivity was 35% (22% to 50%), (moderate‐certainty evidence) and pooled specificity was 95% (89% to 96%), (low‐certainty evidence), (7 studies, 3365 participants, 13% with tuberculosis). For a population of 1000 people where 100 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 80 to be LF‐LAM positive: of these, 45 (56%) would not have tuberculosis (false‐positives); and 920 to be LF‐LAM negative: of these, 65 (7%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 62% (41% to 83%) among inpatients versus 31% (18% to 47%) among outpatients; pooled specificity was 84% (48% to 96%) among inpatients versus 95% (87% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. We found that LF‐LAM has a sensitivity of 42% to diagnose tuberculosis in HIV‐positive individuals with tuberculosis symptoms and 35% in HIV‐positive individuals not assessed for tuberculosis symptoms, consistent with findings reported previously. Regardless of how people are enrolled, sensitivity is higher in inpatients and those with lower CD4 cell, but a concomitant lower specificity. As a simple point‐of‐care test that does not depend upon sputum evaluation, LF‐LAM may assist with the diagnosis of tuberculosis, particularly when a sputum specimen cannot be produced. 17 October 2019 Up to date All studies incorporated from most recent search All studies identified during the most recent search (11 May, 2018) have been incorporated in the review, and no ongoing studies identified.
t136	t136_12	yes	Few studies and participants were included in some analyses and only one study was conducted outside of sub‐Saharan Africa.	The lateral flow urine lipoarabinomannan (LF‐LAM) assay Alere Determine™ TB LAM Ag is recommended by the World Health Organization (WHO) to help detect active tuberculosis in HIV‐positive people with severe HIV disease. This review update asks the question, "does new evidence justify the use of LF‐LAM in a broader group of people?”, and is part of the WHO process for updating guidance on the use of LF‐LAM. Objectives To assess the accuracy of LF‐LAM for the diagnosis of active tuberculosis among HIV‐positive adults with signs and symptoms of tuberculosis (symptomatic participants) and among HIV‐positive adults irrespective of signs and symptoms of tuberculosis (unselected participants not assessed for tuberculosis signs and symptoms). The proposed role for LF‐LAM is as an add on to clinical judgement and with other tests to assist in diagnosing tuberculosis. Search methods We searched the Cochrane Infectious Diseases Group Specialized Register; MEDLINE, Embase, Science Citation Index, Web of Science, Latin American Caribbean Health Sciences Literature, Scopus, the WHO International Clinical Trials Registry Platform, the International Standard Randomized Controlled Trial Number Registry, and ProQuest, without language restriction to 11 May 2018. Selection criteria Randomized trials, cross‐sectional, and observational cohort studies that evaluated LF‐LAM for active tuberculosis (pulmonary and extrapulmonary) in HIV‐positive adults. We included studies that used the manufacturer's recommended threshold for test positivity, either the updated reference card with four bands (grade 1 of 4) or the corresponding prior reference card grade with five bands (grade 2 of 5). The reference standard was culture or nucleic acid amplification test from any body site (microbiological). We considered a higher quality reference standard to be one in which two or more specimen types were evaluated for tuberculosis diagnosis and a lower quality reference standard to be one in which only one specimen type was evaluated. Data collection and analysis Two review authors independently extracted data using a standardized form and REDCap electronic data capture tools. We appraised the quality of studies using the Quality Assessment of Diagnostic Accuracy Studies‐2 (QUADAS‐2) tool and performed meta‐analyses to estimate pooled sensitivity and specificity using a bivariate random‐effects model and a Bayesian approach. We analyzed studies enrolling strictly symptomatic participants separately from those enrolling unselected participants. We investigated pre‐defined sources of heterogeneity including the influence of CD4 count and clinical setting on the accuracy estimates. We assessed the certainty of the evidence using the GRADE approach. We included 15 unique studies (nine new studies and six studies from the original review that met the inclusion criteria): eight studies among symptomatic adults and seven studies among unselected adults. All studies were conducted in low‐ or middle‐income countries. Risk of bias was high in the patient selection and reference standard domains, mainly because studies excluded participants unable to produce sputum and used a lower quality reference standard. Participants with tuberculosis symptoms LF‐LAM pooled sensitivity (95% credible interval (CrI) ) was 42% (31% to 55%) (moderate‐certainty evidence) and pooled specificity was 91% (85% to 95%) (very low‐certainty evidence), (8 studies, 3449 participants, 37% with tuberculosis). For a population of 1000 people where 300 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 189 to be LF‐LAM positive: of these, 63 (33%) would not have tuberculosis (false‐positives); and 811 to be LF‐LAM negative: of these, 174 (21%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 52% (40% to 64%) among inpatients versus 29% (17% to 47%) among outpatients; and pooled specificity was 87% (78% to 93%) among inpatients versus 96% (91% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. Unselected participants not assessed for signs and symptoms of tuberculosis LF‐LAM pooled sensitivity was 35% (22% to 50%), (moderate‐certainty evidence) and pooled specificity was 95% (89% to 96%), (low‐certainty evidence), (7 studies, 3365 participants, 13% with tuberculosis). For a population of 1000 people where 100 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 80 to be LF‐LAM positive: of these, 45 (56%) would not have tuberculosis (false‐positives); and 920 to be LF‐LAM negative: of these, 65 (7%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 62% (41% to 83%) among inpatients versus 31% (18% to 47%) among outpatients; pooled specificity was 84% (48% to 96%) among inpatients versus 95% (87% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. We found that LF‐LAM has a sensitivity of 42% to diagnose tuberculosis in HIV‐positive individuals with tuberculosis symptoms and 35% in HIV‐positive individuals not assessed for tuberculosis symptoms, consistent with findings reported previously. Regardless of how people are enrolled, sensitivity is higher in inpatients and those with lower CD4 cell, but a concomitant lower specificity. As a simple point‐of‐care test that does not depend upon sputum evaluation, LF‐LAM may assist with the diagnosis of tuberculosis, particularly when a sputum specimen cannot be produced. 17 October 2019 Up to date All studies incorporated from most recent search All studies identified during the most recent search (11 May, 2018) have been incorporated in the review, and no ongoing studies identified.
t136	t136_13	no	Among symptomatic participants, in theory, for a population of 1000 people where 300 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 189 to be LF‐LAM positive: of these, 63 (33%) would not have tuberculosis (false‐positives); and 811 to be LF‐LAM negative: of these, 174 (21%) would have tuberculosis (false‐negatives).	The lateral flow urine lipoarabinomannan (LF‐LAM) assay Alere Determine™ TB LAM Ag is recommended by the World Health Organization (WHO) to help detect active tuberculosis in HIV‐positive people with severe HIV disease. This review update asks the question, "does new evidence justify the use of LF‐LAM in a broader group of people?”, and is part of the WHO process for updating guidance on the use of LF‐LAM. Objectives To assess the accuracy of LF‐LAM for the diagnosis of active tuberculosis among HIV‐positive adults with signs and symptoms of tuberculosis (symptomatic participants) and among HIV‐positive adults irrespective of signs and symptoms of tuberculosis (unselected participants not assessed for tuberculosis signs and symptoms). The proposed role for LF‐LAM is as an add on to clinical judgement and with other tests to assist in diagnosing tuberculosis. Search methods We searched the Cochrane Infectious Diseases Group Specialized Register; MEDLINE, Embase, Science Citation Index, Web of Science, Latin American Caribbean Health Sciences Literature, Scopus, the WHO International Clinical Trials Registry Platform, the International Standard Randomized Controlled Trial Number Registry, and ProQuest, without language restriction to 11 May 2018. Selection criteria Randomized trials, cross‐sectional, and observational cohort studies that evaluated LF‐LAM for active tuberculosis (pulmonary and extrapulmonary) in HIV‐positive adults. We included studies that used the manufacturer's recommended threshold for test positivity, either the updated reference card with four bands (grade 1 of 4) or the corresponding prior reference card grade with five bands (grade 2 of 5). The reference standard was culture or nucleic acid amplification test from any body site (microbiological). We considered a higher quality reference standard to be one in which two or more specimen types were evaluated for tuberculosis diagnosis and a lower quality reference standard to be one in which only one specimen type was evaluated. Data collection and analysis Two review authors independently extracted data using a standardized form and REDCap electronic data capture tools. We appraised the quality of studies using the Quality Assessment of Diagnostic Accuracy Studies‐2 (QUADAS‐2) tool and performed meta‐analyses to estimate pooled sensitivity and specificity using a bivariate random‐effects model and a Bayesian approach. We analyzed studies enrolling strictly symptomatic participants separately from those enrolling unselected participants. We investigated pre‐defined sources of heterogeneity including the influence of CD4 count and clinical setting on the accuracy estimates. We assessed the certainty of the evidence using the GRADE approach. We included 15 unique studies (nine new studies and six studies from the original review that met the inclusion criteria): eight studies among symptomatic adults and seven studies among unselected adults. All studies were conducted in low‐ or middle‐income countries. Risk of bias was high in the patient selection and reference standard domains, mainly because studies excluded participants unable to produce sputum and used a lower quality reference standard. Participants with tuberculosis symptoms LF‐LAM pooled sensitivity (95% credible interval (CrI) ) was 42% (31% to 55%) (moderate‐certainty evidence) and pooled specificity was 91% (85% to 95%) (very low‐certainty evidence), (8 studies, 3449 participants, 37% with tuberculosis). For a population of 1000 people where 300 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 189 to be LF‐LAM positive: of these, 63 (33%) would not have tuberculosis (false‐positives); and 811 to be LF‐LAM negative: of these, 174 (21%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 52% (40% to 64%) among inpatients versus 29% (17% to 47%) among outpatients; and pooled specificity was 87% (78% to 93%) among inpatients versus 96% (91% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. Unselected participants not assessed for signs and symptoms of tuberculosis LF‐LAM pooled sensitivity was 35% (22% to 50%), (moderate‐certainty evidence) and pooled specificity was 95% (89% to 96%), (low‐certainty evidence), (7 studies, 3365 participants, 13% with tuberculosis). For a population of 1000 people where 100 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 80 to be LF‐LAM positive: of these, 45 (56%) would not have tuberculosis (false‐positives); and 920 to be LF‐LAM negative: of these, 65 (7%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 62% (41% to 83%) among inpatients versus 31% (18% to 47%) among outpatients; pooled specificity was 84% (48% to 96%) among inpatients versus 95% (87% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. We found that LF‐LAM has a sensitivity of 42% to diagnose tuberculosis in HIV‐positive individuals with tuberculosis symptoms and 35% in HIV‐positive individuals not assessed for tuberculosis symptoms, consistent with findings reported previously. Regardless of how people are enrolled, sensitivity is higher in inpatients and those with lower CD4 cell, but a concomitant lower specificity. As a simple point‐of‐care test that does not depend upon sputum evaluation, LF‐LAM may assist with the diagnosis of tuberculosis, particularly when a sputum specimen cannot be produced. 17 October 2019 Up to date All studies incorporated from most recent search All studies identified during the most recent search (11 May, 2018) have been incorporated in the review, and no ongoing studies identified.
t136	t136_14	no	Among unselected participants, in theory, for a population of 1000 people where 100 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 80 to be LF‐LAM positive: of these, 45 (56%) would not have tuberculosis (false‐positives); and 920 to be LF‐LAM negative: of these, 65 (7%) would have tuberculosis (false‐negatives).	The lateral flow urine lipoarabinomannan (LF‐LAM) assay Alere Determine™ TB LAM Ag is recommended by the World Health Organization (WHO) to help detect active tuberculosis in HIV‐positive people with severe HIV disease. This review update asks the question, "does new evidence justify the use of LF‐LAM in a broader group of people?”, and is part of the WHO process for updating guidance on the use of LF‐LAM. Objectives To assess the accuracy of LF‐LAM for the diagnosis of active tuberculosis among HIV‐positive adults with signs and symptoms of tuberculosis (symptomatic participants) and among HIV‐positive adults irrespective of signs and symptoms of tuberculosis (unselected participants not assessed for tuberculosis signs and symptoms). The proposed role for LF‐LAM is as an add on to clinical judgement and with other tests to assist in diagnosing tuberculosis. Search methods We searched the Cochrane Infectious Diseases Group Specialized Register; MEDLINE, Embase, Science Citation Index, Web of Science, Latin American Caribbean Health Sciences Literature, Scopus, the WHO International Clinical Trials Registry Platform, the International Standard Randomized Controlled Trial Number Registry, and ProQuest, without language restriction to 11 May 2018. Selection criteria Randomized trials, cross‐sectional, and observational cohort studies that evaluated LF‐LAM for active tuberculosis (pulmonary and extrapulmonary) in HIV‐positive adults. We included studies that used the manufacturer's recommended threshold for test positivity, either the updated reference card with four bands (grade 1 of 4) or the corresponding prior reference card grade with five bands (grade 2 of 5). The reference standard was culture or nucleic acid amplification test from any body site (microbiological). We considered a higher quality reference standard to be one in which two or more specimen types were evaluated for tuberculosis diagnosis and a lower quality reference standard to be one in which only one specimen type was evaluated. Data collection and analysis Two review authors independently extracted data using a standardized form and REDCap electronic data capture tools. We appraised the quality of studies using the Quality Assessment of Diagnostic Accuracy Studies‐2 (QUADAS‐2) tool and performed meta‐analyses to estimate pooled sensitivity and specificity using a bivariate random‐effects model and a Bayesian approach. We analyzed studies enrolling strictly symptomatic participants separately from those enrolling unselected participants. We investigated pre‐defined sources of heterogeneity including the influence of CD4 count and clinical setting on the accuracy estimates. We assessed the certainty of the evidence using the GRADE approach. We included 15 unique studies (nine new studies and six studies from the original review that met the inclusion criteria): eight studies among symptomatic adults and seven studies among unselected adults. All studies were conducted in low‐ or middle‐income countries. Risk of bias was high in the patient selection and reference standard domains, mainly because studies excluded participants unable to produce sputum and used a lower quality reference standard. Participants with tuberculosis symptoms LF‐LAM pooled sensitivity (95% credible interval (CrI) ) was 42% (31% to 55%) (moderate‐certainty evidence) and pooled specificity was 91% (85% to 95%) (very low‐certainty evidence), (8 studies, 3449 participants, 37% with tuberculosis). For a population of 1000 people where 300 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 189 to be LF‐LAM positive: of these, 63 (33%) would not have tuberculosis (false‐positives); and 811 to be LF‐LAM negative: of these, 174 (21%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 52% (40% to 64%) among inpatients versus 29% (17% to 47%) among outpatients; and pooled specificity was 87% (78% to 93%) among inpatients versus 96% (91% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. Unselected participants not assessed for signs and symptoms of tuberculosis LF‐LAM pooled sensitivity was 35% (22% to 50%), (moderate‐certainty evidence) and pooled specificity was 95% (89% to 96%), (low‐certainty evidence), (7 studies, 3365 participants, 13% with tuberculosis). For a population of 1000 people where 100 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 80 to be LF‐LAM positive: of these, 45 (56%) would not have tuberculosis (false‐positives); and 920 to be LF‐LAM negative: of these, 65 (7%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 62% (41% to 83%) among inpatients versus 31% (18% to 47%) among outpatients; pooled specificity was 84% (48% to 96%) among inpatients versus 95% (87% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. We found that LF‐LAM has a sensitivity of 42% to diagnose tuberculosis in HIV‐positive individuals with tuberculosis symptoms and 35% in HIV‐positive individuals not assessed for tuberculosis symptoms, consistent with findings reported previously. Regardless of how people are enrolled, sensitivity is higher in inpatients and those with lower CD4 cell, but a concomitant lower specificity. As a simple point‐of‐care test that does not depend upon sputum evaluation, LF‐LAM may assist with the diagnosis of tuberculosis, particularly when a sputum specimen cannot be produced. 17 October 2019 Up to date All studies incorporated from most recent search All studies identified during the most recent search (11 May, 2018) have been incorporated in the review, and no ongoing studies identified.
t136	t136_15	no	HIV‐positive people with tuberculosis symptoms and those not assessed for tuberculosis symptoms.	The lateral flow urine lipoarabinomannan (LF‐LAM) assay Alere Determine™ TB LAM Ag is recommended by the World Health Organization (WHO) to help detect active tuberculosis in HIV‐positive people with severe HIV disease. This review update asks the question, "does new evidence justify the use of LF‐LAM in a broader group of people?”, and is part of the WHO process for updating guidance on the use of LF‐LAM. Objectives To assess the accuracy of LF‐LAM for the diagnosis of active tuberculosis among HIV‐positive adults with signs and symptoms of tuberculosis (symptomatic participants) and among HIV‐positive adults irrespective of signs and symptoms of tuberculosis (unselected participants not assessed for tuberculosis signs and symptoms). The proposed role for LF‐LAM is as an add on to clinical judgement and with other tests to assist in diagnosing tuberculosis. Search methods We searched the Cochrane Infectious Diseases Group Specialized Register; MEDLINE, Embase, Science Citation Index, Web of Science, Latin American Caribbean Health Sciences Literature, Scopus, the WHO International Clinical Trials Registry Platform, the International Standard Randomized Controlled Trial Number Registry, and ProQuest, without language restriction to 11 May 2018. Selection criteria Randomized trials, cross‐sectional, and observational cohort studies that evaluated LF‐LAM for active tuberculosis (pulmonary and extrapulmonary) in HIV‐positive adults. We included studies that used the manufacturer's recommended threshold for test positivity, either the updated reference card with four bands (grade 1 of 4) or the corresponding prior reference card grade with five bands (grade 2 of 5). The reference standard was culture or nucleic acid amplification test from any body site (microbiological). We considered a higher quality reference standard to be one in which two or more specimen types were evaluated for tuberculosis diagnosis and a lower quality reference standard to be one in which only one specimen type was evaluated. Data collection and analysis Two review authors independently extracted data using a standardized form and REDCap electronic data capture tools. We appraised the quality of studies using the Quality Assessment of Diagnostic Accuracy Studies‐2 (QUADAS‐2) tool and performed meta‐analyses to estimate pooled sensitivity and specificity using a bivariate random‐effects model and a Bayesian approach. We analyzed studies enrolling strictly symptomatic participants separately from those enrolling unselected participants. We investigated pre‐defined sources of heterogeneity including the influence of CD4 count and clinical setting on the accuracy estimates. We assessed the certainty of the evidence using the GRADE approach. We included 15 unique studies (nine new studies and six studies from the original review that met the inclusion criteria): eight studies among symptomatic adults and seven studies among unselected adults. All studies were conducted in low‐ or middle‐income countries. Risk of bias was high in the patient selection and reference standard domains, mainly because studies excluded participants unable to produce sputum and used a lower quality reference standard. Participants with tuberculosis symptoms LF‐LAM pooled sensitivity (95% credible interval (CrI) ) was 42% (31% to 55%) (moderate‐certainty evidence) and pooled specificity was 91% (85% to 95%) (very low‐certainty evidence), (8 studies, 3449 participants, 37% with tuberculosis). For a population of 1000 people where 300 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 189 to be LF‐LAM positive: of these, 63 (33%) would not have tuberculosis (false‐positives); and 811 to be LF‐LAM negative: of these, 174 (21%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 52% (40% to 64%) among inpatients versus 29% (17% to 47%) among outpatients; and pooled specificity was 87% (78% to 93%) among inpatients versus 96% (91% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. Unselected participants not assessed for signs and symptoms of tuberculosis LF‐LAM pooled sensitivity was 35% (22% to 50%), (moderate‐certainty evidence) and pooled specificity was 95% (89% to 96%), (low‐certainty evidence), (7 studies, 3365 participants, 13% with tuberculosis). For a population of 1000 people where 100 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 80 to be LF‐LAM positive: of these, 45 (56%) would not have tuberculosis (false‐positives); and 920 to be LF‐LAM negative: of these, 65 (7%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 62% (41% to 83%) among inpatients versus 31% (18% to 47%) among outpatients; pooled specificity was 84% (48% to 96%) among inpatients versus 95% (87% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. We found that LF‐LAM has a sensitivity of 42% to diagnose tuberculosis in HIV‐positive individuals with tuberculosis symptoms and 35% in HIV‐positive individuals not assessed for tuberculosis symptoms, consistent with findings reported previously. Regardless of how people are enrolled, sensitivity is higher in inpatients and those with lower CD4 cell, but a concomitant lower specificity. As a simple point‐of‐care test that does not depend upon sputum evaluation, LF‐LAM may assist with the diagnosis of tuberculosis, particularly when a sputum specimen cannot be produced. 17 October 2019 Up to date All studies incorporated from most recent search All studies identified during the most recent search (11 May, 2018) have been incorporated in the review, and no ongoing studies identified.
t136	t136_16	no	LF‐LAM has sensitivity around 40% to detect tuberculosis.	The lateral flow urine lipoarabinomannan (LF‐LAM) assay Alere Determine™ TB LAM Ag is recommended by the World Health Organization (WHO) to help detect active tuberculosis in HIV‐positive people with severe HIV disease. This review update asks the question, "does new evidence justify the use of LF‐LAM in a broader group of people?”, and is part of the WHO process for updating guidance on the use of LF‐LAM. Objectives To assess the accuracy of LF‐LAM for the diagnosis of active tuberculosis among HIV‐positive adults with signs and symptoms of tuberculosis (symptomatic participants) and among HIV‐positive adults irrespective of signs and symptoms of tuberculosis (unselected participants not assessed for tuberculosis signs and symptoms). The proposed role for LF‐LAM is as an add on to clinical judgement and with other tests to assist in diagnosing tuberculosis. Search methods We searched the Cochrane Infectious Diseases Group Specialized Register; MEDLINE, Embase, Science Citation Index, Web of Science, Latin American Caribbean Health Sciences Literature, Scopus, the WHO International Clinical Trials Registry Platform, the International Standard Randomized Controlled Trial Number Registry, and ProQuest, without language restriction to 11 May 2018. Selection criteria Randomized trials, cross‐sectional, and observational cohort studies that evaluated LF‐LAM for active tuberculosis (pulmonary and extrapulmonary) in HIV‐positive adults. We included studies that used the manufacturer's recommended threshold for test positivity, either the updated reference card with four bands (grade 1 of 4) or the corresponding prior reference card grade with five bands (grade 2 of 5). The reference standard was culture or nucleic acid amplification test from any body site (microbiological). We considered a higher quality reference standard to be one in which two or more specimen types were evaluated for tuberculosis diagnosis and a lower quality reference standard to be one in which only one specimen type was evaluated. Data collection and analysis Two review authors independently extracted data using a standardized form and REDCap electronic data capture tools. We appraised the quality of studies using the Quality Assessment of Diagnostic Accuracy Studies‐2 (QUADAS‐2) tool and performed meta‐analyses to estimate pooled sensitivity and specificity using a bivariate random‐effects model and a Bayesian approach. We analyzed studies enrolling strictly symptomatic participants separately from those enrolling unselected participants. We investigated pre‐defined sources of heterogeneity including the influence of CD4 count and clinical setting on the accuracy estimates. We assessed the certainty of the evidence using the GRADE approach. We included 15 unique studies (nine new studies and six studies from the original review that met the inclusion criteria): eight studies among symptomatic adults and seven studies among unselected adults. All studies were conducted in low‐ or middle‐income countries. Risk of bias was high in the patient selection and reference standard domains, mainly because studies excluded participants unable to produce sputum and used a lower quality reference standard. Participants with tuberculosis symptoms LF‐LAM pooled sensitivity (95% credible interval (CrI) ) was 42% (31% to 55%) (moderate‐certainty evidence) and pooled specificity was 91% (85% to 95%) (very low‐certainty evidence), (8 studies, 3449 participants, 37% with tuberculosis). For a population of 1000 people where 300 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 189 to be LF‐LAM positive: of these, 63 (33%) would not have tuberculosis (false‐positives); and 811 to be LF‐LAM negative: of these, 174 (21%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 52% (40% to 64%) among inpatients versus 29% (17% to 47%) among outpatients; and pooled specificity was 87% (78% to 93%) among inpatients versus 96% (91% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. Unselected participants not assessed for signs and symptoms of tuberculosis LF‐LAM pooled sensitivity was 35% (22% to 50%), (moderate‐certainty evidence) and pooled specificity was 95% (89% to 96%), (low‐certainty evidence), (7 studies, 3365 participants, 13% with tuberculosis). For a population of 1000 people where 100 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 80 to be LF‐LAM positive: of these, 45 (56%) would not have tuberculosis (false‐positives); and 920 to be LF‐LAM negative: of these, 65 (7%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 62% (41% to 83%) among inpatients versus 31% (18% to 47%) among outpatients; pooled specificity was 84% (48% to 96%) among inpatients versus 95% (87% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. We found that LF‐LAM has a sensitivity of 42% to diagnose tuberculosis in HIV‐positive individuals with tuberculosis symptoms and 35% in HIV‐positive individuals not assessed for tuberculosis symptoms, consistent with findings reported previously. Regardless of how people are enrolled, sensitivity is higher in inpatients and those with lower CD4 cell, but a concomitant lower specificity. As a simple point‐of‐care test that does not depend upon sputum evaluation, LF‐LAM may assist with the diagnosis of tuberculosis, particularly when a sputum specimen cannot be produced. 17 October 2019 Up to date All studies incorporated from most recent search All studies identified during the most recent search (11 May, 2018) have been incorporated in the review, and no ongoing studies identified.
t136	t136_17	yes	As the test does not require sputum collection, LF‐LAM may be the only way to diagnose tuberculosis when sputum cannot be produced.	The lateral flow urine lipoarabinomannan (LF‐LAM) assay Alere Determine™ TB LAM Ag is recommended by the World Health Organization (WHO) to help detect active tuberculosis in HIV‐positive people with severe HIV disease. This review update asks the question, "does new evidence justify the use of LF‐LAM in a broader group of people?”, and is part of the WHO process for updating guidance on the use of LF‐LAM. Objectives To assess the accuracy of LF‐LAM for the diagnosis of active tuberculosis among HIV‐positive adults with signs and symptoms of tuberculosis (symptomatic participants) and among HIV‐positive adults irrespective of signs and symptoms of tuberculosis (unselected participants not assessed for tuberculosis signs and symptoms). The proposed role for LF‐LAM is as an add on to clinical judgement and with other tests to assist in diagnosing tuberculosis. Search methods We searched the Cochrane Infectious Diseases Group Specialized Register; MEDLINE, Embase, Science Citation Index, Web of Science, Latin American Caribbean Health Sciences Literature, Scopus, the WHO International Clinical Trials Registry Platform, the International Standard Randomized Controlled Trial Number Registry, and ProQuest, without language restriction to 11 May 2018. Selection criteria Randomized trials, cross‐sectional, and observational cohort studies that evaluated LF‐LAM for active tuberculosis (pulmonary and extrapulmonary) in HIV‐positive adults. We included studies that used the manufacturer's recommended threshold for test positivity, either the updated reference card with four bands (grade 1 of 4) or the corresponding prior reference card grade with five bands (grade 2 of 5). The reference standard was culture or nucleic acid amplification test from any body site (microbiological). We considered a higher quality reference standard to be one in which two or more specimen types were evaluated for tuberculosis diagnosis and a lower quality reference standard to be one in which only one specimen type was evaluated. Data collection and analysis Two review authors independently extracted data using a standardized form and REDCap electronic data capture tools. We appraised the quality of studies using the Quality Assessment of Diagnostic Accuracy Studies‐2 (QUADAS‐2) tool and performed meta‐analyses to estimate pooled sensitivity and specificity using a bivariate random‐effects model and a Bayesian approach. We analyzed studies enrolling strictly symptomatic participants separately from those enrolling unselected participants. We investigated pre‐defined sources of heterogeneity including the influence of CD4 count and clinical setting on the accuracy estimates. We assessed the certainty of the evidence using the GRADE approach. We included 15 unique studies (nine new studies and six studies from the original review that met the inclusion criteria): eight studies among symptomatic adults and seven studies among unselected adults. All studies were conducted in low‐ or middle‐income countries. Risk of bias was high in the patient selection and reference standard domains, mainly because studies excluded participants unable to produce sputum and used a lower quality reference standard. Participants with tuberculosis symptoms LF‐LAM pooled sensitivity (95% credible interval (CrI) ) was 42% (31% to 55%) (moderate‐certainty evidence) and pooled specificity was 91% (85% to 95%) (very low‐certainty evidence), (8 studies, 3449 participants, 37% with tuberculosis). For a population of 1000 people where 300 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 189 to be LF‐LAM positive: of these, 63 (33%) would not have tuberculosis (false‐positives); and 811 to be LF‐LAM negative: of these, 174 (21%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 52% (40% to 64%) among inpatients versus 29% (17% to 47%) among outpatients; and pooled specificity was 87% (78% to 93%) among inpatients versus 96% (91% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. Unselected participants not assessed for signs and symptoms of tuberculosis LF‐LAM pooled sensitivity was 35% (22% to 50%), (moderate‐certainty evidence) and pooled specificity was 95% (89% to 96%), (low‐certainty evidence), (7 studies, 3365 participants, 13% with tuberculosis). For a population of 1000 people where 100 have microbiologically‐confirmed tuberculosis, the utilization of LF‐LAM would result in: 80 to be LF‐LAM positive: of these, 45 (56%) would not have tuberculosis (false‐positives); and 920 to be LF‐LAM negative: of these, 65 (7%) would have tuberculosis (false‐negatives). By clinical setting, pooled sensitivity was 62% (41% to 83%) among inpatients versus 31% (18% to 47%) among outpatients; pooled specificity was 84% (48% to 96%) among inpatients versus 95% (87% to 99%) among outpatients. Stratified by CD4 cell count, pooled sensitivity increased, and specificity decreased with lower CD4 cell count. We found that LF‐LAM has a sensitivity of 42% to diagnose tuberculosis in HIV‐positive individuals with tuberculosis symptoms and 35% in HIV‐positive individuals not assessed for tuberculosis symptoms, consistent with findings reported previously. Regardless of how people are enrolled, sensitivity is higher in inpatients and those with lower CD4 cell, but a concomitant lower specificity. As a simple point‐of‐care test that does not depend upon sputum evaluation, LF‐LAM may assist with the diagnosis of tuberculosis, particularly when a sputum specimen cannot be produced. 17 October 2019 Up to date All studies incorporated from most recent search All studies identified during the most recent search (11 May, 2018) have been incorporated in the review, and no ongoing studies identified.
t137	t137_1	no	Melatonin is widely used for management of sleep disorders in children with poor or no vision.	Exogenous melatonin helps in regulating the circadian rhythm and is widely used for the management of sleep disorders in visually impaired children. Objectives The aim of the review was to assess melatonin therapy for treatment of non‐respiratory sleep disorders in visually impaired children, with regard to improvement in sleep habit, sleep scheduling and sleep maintenance, when compared with placebo or no treatment. Search methods We searched the following databases between February 2011 and July 2011: the Cochrane Central Register of Controlled Trials (CENTRAL) 2011(1) searched on 4th February 2011; MEDLINE (1950 to June Week 3, 2011) searched on 20th June 2011; EMBASE (1980 to June Week 4, 2011) searched on 7th July 2011; CINAHL (1937 to 21 September 2011); the metaRegister of Controlled Trials (this includes ClinicalTrial.gov) searched 20 July 2011, and reference lists of papers identified after initial screening. Selection criteria We planned to include randomized controlled trials (RCTs) and quasi‐RCTs, including cross‐over studies. Treatment would be exogenous melatonin. Control groups could be placebo, other medication for sleep disorders or no treatment. Outcomes sought were improved sleep with regard to timing and duration, quality of life and adverse events. Data collection and analysis Three review authors independently assessed trials for inclusion in the review. We did not find any studies fulfilling the inclusion criteria, therefore no outcome data are reported. We identified nine studies after initial screening and, after further evaluation, we excluded these. The excluded studies involved a total of 163 individuals aged two years to 18 years. We excluded studies for three main reasons: they were non‐randomized or case series studies, they were studies of people over 18 years of age or even where the study was randomised, the study population was mixed and results pertaining to the visually impaired cohort could not be independently evaluated. No significant adverse effects of melatonin were reported in these excluded studies. There is currently no high quality data to support or refute the use of melatonin for sleep disorders in visually impaired children. Placebo‐controlled trials examining important clinical outcomes such as sleep quality, sleep latency, duration of sleep and night‐time awakenings are needed. As the numbers of children meeting study inclusion criteria are likely to be low at individual sites, multicentre collaboration between developmental paediatricians, sleep physicians and other health care professionals is essential to achieve sufficient sample size for controlled studies. Such collaboration would help facilitate local recruitment at multiple sites, with study oversight being provided by paediatricians with expertise in sleep disorders. Participation of collaborators with experience in evidence‐based practice research is also desirable due to the lack of protocols on melatonin therapy in the target population.
t137	t137_2	no	The current review planned to examine studies on the use of melatonin in these children to determine whether this drug is effective for improving their sleep (safety is not mentioned in objectives or abstract and adverse effects is a secondary outcome).	Exogenous melatonin helps in regulating the circadian rhythm and is widely used for the management of sleep disorders in visually impaired children. Objectives The aim of the review was to assess melatonin therapy for treatment of non‐respiratory sleep disorders in visually impaired children, with regard to improvement in sleep habit, sleep scheduling and sleep maintenance, when compared with placebo or no treatment. Search methods We searched the following databases between February 2011 and July 2011: the Cochrane Central Register of Controlled Trials (CENTRAL) 2011(1) searched on 4th February 2011; MEDLINE (1950 to June Week 3, 2011) searched on 20th June 2011; EMBASE (1980 to June Week 4, 2011) searched on 7th July 2011; CINAHL (1937 to 21 September 2011); the metaRegister of Controlled Trials (this includes ClinicalTrial.gov) searched 20 July 2011, and reference lists of papers identified after initial screening. Selection criteria We planned to include randomized controlled trials (RCTs) and quasi‐RCTs, including cross‐over studies. Treatment would be exogenous melatonin. Control groups could be placebo, other medication for sleep disorders or no treatment. Outcomes sought were improved sleep with regard to timing and duration, quality of life and adverse events. Data collection and analysis Three review authors independently assessed trials for inclusion in the review. We did not find any studies fulfilling the inclusion criteria, therefore no outcome data are reported. We identified nine studies after initial screening and, after further evaluation, we excluded these. The excluded studies involved a total of 163 individuals aged two years to 18 years. We excluded studies for three main reasons: they were non‐randomized or case series studies, they were studies of people over 18 years of age or even where the study was randomised, the study population was mixed and results pertaining to the visually impaired cohort could not be independently evaluated. No significant adverse effects of melatonin were reported in these excluded studies. There is currently no high quality data to support or refute the use of melatonin for sleep disorders in visually impaired children. Placebo‐controlled trials examining important clinical outcomes such as sleep quality, sleep latency, duration of sleep and night‐time awakenings are needed. As the numbers of children meeting study inclusion criteria are likely to be low at individual sites, multicentre collaboration between developmental paediatricians, sleep physicians and other health care professionals is essential to achieve sufficient sample size for controlled studies. Such collaboration would help facilitate local recruitment at multiple sites, with study oversight being provided by paediatricians with expertise in sleep disorders. Participation of collaborators with experience in evidence‐based practice research is also desirable due to the lack of protocols on melatonin therapy in the target population.
t137	t137_3	no	We only wanted to use studies where the children had been randomly allocated to a treatment group and a control group that got no treatment or another medication or a placebo.	Exogenous melatonin helps in regulating the circadian rhythm and is widely used for the management of sleep disorders in visually impaired children. Objectives The aim of the review was to assess melatonin therapy for treatment of non‐respiratory sleep disorders in visually impaired children, with regard to improvement in sleep habit, sleep scheduling and sleep maintenance, when compared with placebo or no treatment. Search methods We searched the following databases between February 2011 and July 2011: the Cochrane Central Register of Controlled Trials (CENTRAL) 2011(1) searched on 4th February 2011; MEDLINE (1950 to June Week 3, 2011) searched on 20th June 2011; EMBASE (1980 to June Week 4, 2011) searched on 7th July 2011; CINAHL (1937 to 21 September 2011); the metaRegister of Controlled Trials (this includes ClinicalTrial.gov) searched 20 July 2011, and reference lists of papers identified after initial screening. Selection criteria We planned to include randomized controlled trials (RCTs) and quasi‐RCTs, including cross‐over studies. Treatment would be exogenous melatonin. Control groups could be placebo, other medication for sleep disorders or no treatment. Outcomes sought were improved sleep with regard to timing and duration, quality of life and adverse events. Data collection and analysis Three review authors independently assessed trials for inclusion in the review. We did not find any studies fulfilling the inclusion criteria, therefore no outcome data are reported. We identified nine studies after initial screening and, after further evaluation, we excluded these. The excluded studies involved a total of 163 individuals aged two years to 18 years. We excluded studies for three main reasons: they were non‐randomized or case series studies, they were studies of people over 18 years of age or even where the study was randomised, the study population was mixed and results pertaining to the visually impaired cohort could not be independently evaluated. No significant adverse effects of melatonin were reported in these excluded studies. There is currently no high quality data to support or refute the use of melatonin for sleep disorders in visually impaired children. Placebo‐controlled trials examining important clinical outcomes such as sleep quality, sleep latency, duration of sleep and night‐time awakenings are needed. As the numbers of children meeting study inclusion criteria are likely to be low at individual sites, multicentre collaboration between developmental paediatricians, sleep physicians and other health care professionals is essential to achieve sufficient sample size for controlled studies. Such collaboration would help facilitate local recruitment at multiple sites, with study oversight being provided by paediatricians with expertise in sleep disorders. Participation of collaborators with experience in evidence‐based practice research is also desirable due to the lack of protocols on melatonin therapy in the target population.
t137	t137_4	no	We did not find any of these studies that were suitable to be included in our review and so we are unable to draw any conclusions about whether or not melatonin improves sleep for visually impaired children.	Exogenous melatonin helps in regulating the circadian rhythm and is widely used for the management of sleep disorders in visually impaired children. Objectives The aim of the review was to assess melatonin therapy for treatment of non‐respiratory sleep disorders in visually impaired children, with regard to improvement in sleep habit, sleep scheduling and sleep maintenance, when compared with placebo or no treatment. Search methods We searched the following databases between February 2011 and July 2011: the Cochrane Central Register of Controlled Trials (CENTRAL) 2011(1) searched on 4th February 2011; MEDLINE (1950 to June Week 3, 2011) searched on 20th June 2011; EMBASE (1980 to June Week 4, 2011) searched on 7th July 2011; CINAHL (1937 to 21 September 2011); the metaRegister of Controlled Trials (this includes ClinicalTrial.gov) searched 20 July 2011, and reference lists of papers identified after initial screening. Selection criteria We planned to include randomized controlled trials (RCTs) and quasi‐RCTs, including cross‐over studies. Treatment would be exogenous melatonin. Control groups could be placebo, other medication for sleep disorders or no treatment. Outcomes sought were improved sleep with regard to timing and duration, quality of life and adverse events. Data collection and analysis Three review authors independently assessed trials for inclusion in the review. We did not find any studies fulfilling the inclusion criteria, therefore no outcome data are reported. We identified nine studies after initial screening and, after further evaluation, we excluded these. The excluded studies involved a total of 163 individuals aged two years to 18 years. We excluded studies for three main reasons: they were non‐randomized or case series studies, they were studies of people over 18 years of age or even where the study was randomised, the study population was mixed and results pertaining to the visually impaired cohort could not be independently evaluated. No significant adverse effects of melatonin were reported in these excluded studies. There is currently no high quality data to support or refute the use of melatonin for sleep disorders in visually impaired children. Placebo‐controlled trials examining important clinical outcomes such as sleep quality, sleep latency, duration of sleep and night‐time awakenings are needed. As the numbers of children meeting study inclusion criteria are likely to be low at individual sites, multicentre collaboration between developmental paediatricians, sleep physicians and other health care professionals is essential to achieve sufficient sample size for controlled studies. Such collaboration would help facilitate local recruitment at multiple sites, with study oversight being provided by paediatricians with expertise in sleep disorders. Participation of collaborators with experience in evidence‐based practice research is also desirable due to the lack of protocols on melatonin therapy in the target population.
t137	t137_5	no	To find out, we need appropriately designed clinical trials.	Exogenous melatonin helps in regulating the circadian rhythm and is widely used for the management of sleep disorders in visually impaired children. Objectives The aim of the review was to assess melatonin therapy for treatment of non‐respiratory sleep disorders in visually impaired children, with regard to improvement in sleep habit, sleep scheduling and sleep maintenance, when compared with placebo or no treatment. Search methods We searched the following databases between February 2011 and July 2011: the Cochrane Central Register of Controlled Trials (CENTRAL) 2011(1) searched on 4th February 2011; MEDLINE (1950 to June Week 3, 2011) searched on 20th June 2011; EMBASE (1980 to June Week 4, 2011) searched on 7th July 2011; CINAHL (1937 to 21 September 2011); the metaRegister of Controlled Trials (this includes ClinicalTrial.gov) searched 20 July 2011, and reference lists of papers identified after initial screening. Selection criteria We planned to include randomized controlled trials (RCTs) and quasi‐RCTs, including cross‐over studies. Treatment would be exogenous melatonin. Control groups could be placebo, other medication for sleep disorders or no treatment. Outcomes sought were improved sleep with regard to timing and duration, quality of life and adverse events. Data collection and analysis Three review authors independently assessed trials for inclusion in the review. We did not find any studies fulfilling the inclusion criteria, therefore no outcome data are reported. We identified nine studies after initial screening and, after further evaluation, we excluded these. The excluded studies involved a total of 163 individuals aged two years to 18 years. We excluded studies for three main reasons: they were non‐randomized or case series studies, they were studies of people over 18 years of age or even where the study was randomised, the study population was mixed and results pertaining to the visually impaired cohort could not be independently evaluated. No significant adverse effects of melatonin were reported in these excluded studies. There is currently no high quality data to support or refute the use of melatonin for sleep disorders in visually impaired children. Placebo‐controlled trials examining important clinical outcomes such as sleep quality, sleep latency, duration of sleep and night‐time awakenings are needed. As the numbers of children meeting study inclusion criteria are likely to be low at individual sites, multicentre collaboration between developmental paediatricians, sleep physicians and other health care professionals is essential to achieve sufficient sample size for controlled studies. Such collaboration would help facilitate local recruitment at multiple sites, with study oversight being provided by paediatricians with expertise in sleep disorders. Participation of collaborators with experience in evidence‐based practice research is also desirable due to the lack of protocols on melatonin therapy in the target population.
t137	t137_6	no	Due to lack of knowledge about best practice in the use of melatonin with these children, it would be useful to have researchers involved who are experienced in sleep disorders and in evidence‐based practice research.	Exogenous melatonin helps in regulating the circadian rhythm and is widely used for the management of sleep disorders in visually impaired children. Objectives The aim of the review was to assess melatonin therapy for treatment of non‐respiratory sleep disorders in visually impaired children, with regard to improvement in sleep habit, sleep scheduling and sleep maintenance, when compared with placebo or no treatment. Search methods We searched the following databases between February 2011 and July 2011: the Cochrane Central Register of Controlled Trials (CENTRAL) 2011(1) searched on 4th February 2011; MEDLINE (1950 to June Week 3, 2011) searched on 20th June 2011; EMBASE (1980 to June Week 4, 2011) searched on 7th July 2011; CINAHL (1937 to 21 September 2011); the metaRegister of Controlled Trials (this includes ClinicalTrial.gov) searched 20 July 2011, and reference lists of papers identified after initial screening. Selection criteria We planned to include randomized controlled trials (RCTs) and quasi‐RCTs, including cross‐over studies. Treatment would be exogenous melatonin. Control groups could be placebo, other medication for sleep disorders or no treatment. Outcomes sought were improved sleep with regard to timing and duration, quality of life and adverse events. Data collection and analysis Three review authors independently assessed trials for inclusion in the review. We did not find any studies fulfilling the inclusion criteria, therefore no outcome data are reported. We identified nine studies after initial screening and, after further evaluation, we excluded these. The excluded studies involved a total of 163 individuals aged two years to 18 years. We excluded studies for three main reasons: they were non‐randomized or case series studies, they were studies of people over 18 years of age or even where the study was randomised, the study population was mixed and results pertaining to the visually impaired cohort could not be independently evaluated. No significant adverse effects of melatonin were reported in these excluded studies. There is currently no high quality data to support or refute the use of melatonin for sleep disorders in visually impaired children. Placebo‐controlled trials examining important clinical outcomes such as sleep quality, sleep latency, duration of sleep and night‐time awakenings are needed. As the numbers of children meeting study inclusion criteria are likely to be low at individual sites, multicentre collaboration between developmental paediatricians, sleep physicians and other health care professionals is essential to achieve sufficient sample size for controlled studies. Such collaboration would help facilitate local recruitment at multiple sites, with study oversight being provided by paediatricians with expertise in sleep disorders. Participation of collaborators with experience in evidence‐based practice research is also desirable due to the lack of protocols on melatonin therapy in the target population.
t137	t137_7	yes	In addition, studies involving more than one location would be beneficial to increase the number of children being evaluated and make it more likely we will reach solid conclusions about whether melatonin works for this group of children, as well as details about the most effective dosage and timing of the treatment.	Exogenous melatonin helps in regulating the circadian rhythm and is widely used for the management of sleep disorders in visually impaired children. Objectives The aim of the review was to assess melatonin therapy for treatment of non‐respiratory sleep disorders in visually impaired children, with regard to improvement in sleep habit, sleep scheduling and sleep maintenance, when compared with placebo or no treatment. Search methods We searched the following databases between February 2011 and July 2011: the Cochrane Central Register of Controlled Trials (CENTRAL) 2011(1) searched on 4th February 2011; MEDLINE (1950 to June Week 3, 2011) searched on 20th June 2011; EMBASE (1980 to June Week 4, 2011) searched on 7th July 2011; CINAHL (1937 to 21 September 2011); the metaRegister of Controlled Trials (this includes ClinicalTrial.gov) searched 20 July 2011, and reference lists of papers identified after initial screening. Selection criteria We planned to include randomized controlled trials (RCTs) and quasi‐RCTs, including cross‐over studies. Treatment would be exogenous melatonin. Control groups could be placebo, other medication for sleep disorders or no treatment. Outcomes sought were improved sleep with regard to timing and duration, quality of life and adverse events. Data collection and analysis Three review authors independently assessed trials for inclusion in the review. We did not find any studies fulfilling the inclusion criteria, therefore no outcome data are reported. We identified nine studies after initial screening and, after further evaluation, we excluded these. The excluded studies involved a total of 163 individuals aged two years to 18 years. We excluded studies for three main reasons: they were non‐randomized or case series studies, they were studies of people over 18 years of age or even where the study was randomised, the study population was mixed and results pertaining to the visually impaired cohort could not be independently evaluated. No significant adverse effects of melatonin were reported in these excluded studies. There is currently no high quality data to support or refute the use of melatonin for sleep disorders in visually impaired children. Placebo‐controlled trials examining important clinical outcomes such as sleep quality, sleep latency, duration of sleep and night‐time awakenings are needed. As the numbers of children meeting study inclusion criteria are likely to be low at individual sites, multicentre collaboration between developmental paediatricians, sleep physicians and other health care professionals is essential to achieve sufficient sample size for controlled studies. Such collaboration would help facilitate local recruitment at multiple sites, with study oversight being provided by paediatricians with expertise in sleep disorders. Participation of collaborators with experience in evidence‐based practice research is also desirable due to the lack of protocols on melatonin therapy in the target population.
t138	t138_1	no	Toxic epidermal necrolysis (TEN or Lyell's disease) is a rare life‐threatening skin condition.	Toxic epidermal necrolysis is a rare condition where a drug reaction induces skin loss, similar to that seen in extensive burns. It is associated with high morbidity and mortality and there is no clear agreement on effective treatment. Objectives To assess the effects of all interventions for the treatment of toxic epidermal necrolysis. Search methods We searched the Cochrane Skin Group Specialised Register (March 2001), the Cochrane Controlled Trials Register (March 2001), MEDLINE (1966 to December 2001), EMBASE (1980 to December 2001), DARE (4th Quarter 2001) and CINAHL (1982 to October 2001). Selection criteria Randomised controlled trials of therapeutic and supportive interventions that included participants clinically diagnosed with toxic epidermal necrolysis were included. Data collection and analysis Two independent authors carried out study selection and assessment of methodological quality. Only one randomised controlled trial of treatment was identified. This trial compared the effectiveness of thalidomide with placebo and included 22 patients, 12 in the treatment group and 10 in the placebo group. Patients on the treatment arm received thalidomide 200 mg twice daily for 5 days. The main end point was the measurement of the progression of skin detachment after seven days. Other end points were the overall mortality and severity of the disease evaluated with the simplified acute physiology score. The study was terminated as the mortality on the treatment arm was 83% compared to 30% on the control arm (relative risk 2.78, 95% confidence interval 1.04 to 7.40). No randomised controlled trials of the most commonly used current treatments i.e. systemic steroids, cyclosporin A and intravenous immunoglobulins were found. Treatment with thalidomide was not shown to be effective and was associated with significantly higher mortality than placebo. There is no reliable evidence on which to base treatment for toxic epidermal necrolysis, a disease commonly associated with mortality rates of around 30%. More research is required to understand the mechanisms of toxic epidermal necrolysis. International multi‐centre studies are needed in the form of randomised controlled trials, to evaluate treatments for toxic epidermal necrolysis, especially those using high doses of steroid and intravenous immunoglobulins.
t138	t138_2	yes	It is probably an immune response triggered by some drugs or infection, which is more likely to happen in people with suppressed immunity.	Toxic epidermal necrolysis is a rare condition where a drug reaction induces skin loss, similar to that seen in extensive burns. It is associated with high morbidity and mortality and there is no clear agreement on effective treatment. Objectives To assess the effects of all interventions for the treatment of toxic epidermal necrolysis. Search methods We searched the Cochrane Skin Group Specialised Register (March 2001), the Cochrane Controlled Trials Register (March 2001), MEDLINE (1966 to December 2001), EMBASE (1980 to December 2001), DARE (4th Quarter 2001) and CINAHL (1982 to October 2001). Selection criteria Randomised controlled trials of therapeutic and supportive interventions that included participants clinically diagnosed with toxic epidermal necrolysis were included. Data collection and analysis Two independent authors carried out study selection and assessment of methodological quality. Only one randomised controlled trial of treatment was identified. This trial compared the effectiveness of thalidomide with placebo and included 22 patients, 12 in the treatment group and 10 in the placebo group. Patients on the treatment arm received thalidomide 200 mg twice daily for 5 days. The main end point was the measurement of the progression of skin detachment after seven days. Other end points were the overall mortality and severity of the disease evaluated with the simplified acute physiology score. The study was terminated as the mortality on the treatment arm was 83% compared to 30% on the control arm (relative risk 2.78, 95% confidence interval 1.04 to 7.40). No randomised controlled trials of the most commonly used current treatments i.e. systemic steroids, cyclosporin A and intravenous immunoglobulins were found. Treatment with thalidomide was not shown to be effective and was associated with significantly higher mortality than placebo. There is no reliable evidence on which to base treatment for toxic epidermal necrolysis, a disease commonly associated with mortality rates of around 30%. More research is required to understand the mechanisms of toxic epidermal necrolysis. International multi‐centre studies are needed in the form of randomised controlled trials, to evaluate treatments for toxic epidermal necrolysis, especially those using high doses of steroid and intravenous immunoglobulins.
t138	t138_3	yes	TEN causes extensive blistering and shedding of skin, similar to burns.	Toxic epidermal necrolysis is a rare condition where a drug reaction induces skin loss, similar to that seen in extensive burns. It is associated with high morbidity and mortality and there is no clear agreement on effective treatment. Objectives To assess the effects of all interventions for the treatment of toxic epidermal necrolysis. Search methods We searched the Cochrane Skin Group Specialised Register (March 2001), the Cochrane Controlled Trials Register (March 2001), MEDLINE (1966 to December 2001), EMBASE (1980 to December 2001), DARE (4th Quarter 2001) and CINAHL (1982 to October 2001). Selection criteria Randomised controlled trials of therapeutic and supportive interventions that included participants clinically diagnosed with toxic epidermal necrolysis were included. Data collection and analysis Two independent authors carried out study selection and assessment of methodological quality. Only one randomised controlled trial of treatment was identified. This trial compared the effectiveness of thalidomide with placebo and included 22 patients, 12 in the treatment group and 10 in the placebo group. Patients on the treatment arm received thalidomide 200 mg twice daily for 5 days. The main end point was the measurement of the progression of skin detachment after seven days. Other end points were the overall mortality and severity of the disease evaluated with the simplified acute physiology score. The study was terminated as the mortality on the treatment arm was 83% compared to 30% on the control arm (relative risk 2.78, 95% confidence interval 1.04 to 7.40). No randomised controlled trials of the most commonly used current treatments i.e. systemic steroids, cyclosporin A and intravenous immunoglobulins were found. Treatment with thalidomide was not shown to be effective and was associated with significantly higher mortality than placebo. There is no reliable evidence on which to base treatment for toxic epidermal necrolysis, a disease commonly associated with mortality rates of around 30%. More research is required to understand the mechanisms of toxic epidermal necrolysis. International multi‐centre studies are needed in the form of randomised controlled trials, to evaluate treatments for toxic epidermal necrolysis, especially those using high doses of steroid and intravenous immunoglobulins.
t138	t138_4	yes	Drugs used include oral steroids, thalidomide, immunosuppressants and immunoglobulins.	Toxic epidermal necrolysis is a rare condition where a drug reaction induces skin loss, similar to that seen in extensive burns. It is associated with high morbidity and mortality and there is no clear agreement on effective treatment. Objectives To assess the effects of all interventions for the treatment of toxic epidermal necrolysis. Search methods We searched the Cochrane Skin Group Specialised Register (March 2001), the Cochrane Controlled Trials Register (March 2001), MEDLINE (1966 to December 2001), EMBASE (1980 to December 2001), DARE (4th Quarter 2001) and CINAHL (1982 to October 2001). Selection criteria Randomised controlled trials of therapeutic and supportive interventions that included participants clinically diagnosed with toxic epidermal necrolysis were included. Data collection and analysis Two independent authors carried out study selection and assessment of methodological quality. Only one randomised controlled trial of treatment was identified. This trial compared the effectiveness of thalidomide with placebo and included 22 patients, 12 in the treatment group and 10 in the placebo group. Patients on the treatment arm received thalidomide 200 mg twice daily for 5 days. The main end point was the measurement of the progression of skin detachment after seven days. Other end points were the overall mortality and severity of the disease evaluated with the simplified acute physiology score. The study was terminated as the mortality on the treatment arm was 83% compared to 30% on the control arm (relative risk 2.78, 95% confidence interval 1.04 to 7.40). No randomised controlled trials of the most commonly used current treatments i.e. systemic steroids, cyclosporin A and intravenous immunoglobulins were found. Treatment with thalidomide was not shown to be effective and was associated with significantly higher mortality than placebo. There is no reliable evidence on which to base treatment for toxic epidermal necrolysis, a disease commonly associated with mortality rates of around 30%. More research is required to understand the mechanisms of toxic epidermal necrolysis. International multi‐centre studies are needed in the form of randomised controlled trials, to evaluate treatments for toxic epidermal necrolysis, especially those using high doses of steroid and intravenous immunoglobulins.
t138	t138_5	no	This review of trials did not find any reliable evidence for the treatment of TEN.	Toxic epidermal necrolysis is a rare condition where a drug reaction induces skin loss, similar to that seen in extensive burns. It is associated with high morbidity and mortality and there is no clear agreement on effective treatment. Objectives To assess the effects of all interventions for the treatment of toxic epidermal necrolysis. Search methods We searched the Cochrane Skin Group Specialised Register (March 2001), the Cochrane Controlled Trials Register (March 2001), MEDLINE (1966 to December 2001), EMBASE (1980 to December 2001), DARE (4th Quarter 2001) and CINAHL (1982 to October 2001). Selection criteria Randomised controlled trials of therapeutic and supportive interventions that included participants clinically diagnosed with toxic epidermal necrolysis were included. Data collection and analysis Two independent authors carried out study selection and assessment of methodological quality. Only one randomised controlled trial of treatment was identified. This trial compared the effectiveness of thalidomide with placebo and included 22 patients, 12 in the treatment group and 10 in the placebo group. Patients on the treatment arm received thalidomide 200 mg twice daily for 5 days. The main end point was the measurement of the progression of skin detachment after seven days. Other end points were the overall mortality and severity of the disease evaluated with the simplified acute physiology score. The study was terminated as the mortality on the treatment arm was 83% compared to 30% on the control arm (relative risk 2.78, 95% confidence interval 1.04 to 7.40). No randomised controlled trials of the most commonly used current treatments i.e. systemic steroids, cyclosporin A and intravenous immunoglobulins were found. Treatment with thalidomide was not shown to be effective and was associated with significantly higher mortality than placebo. There is no reliable evidence on which to base treatment for toxic epidermal necrolysis, a disease commonly associated with mortality rates of around 30%. More research is required to understand the mechanisms of toxic epidermal necrolysis. International multi‐centre studies are needed in the form of randomised controlled trials, to evaluate treatments for toxic epidermal necrolysis, especially those using high doses of steroid and intravenous immunoglobulins.
t138	t138_6	no	The only trial available used thalidomide, but this trial did not show any benefit from treatment compared against placebo but highlighted increased chances of dying from the treatment.	Toxic epidermal necrolysis is a rare condition where a drug reaction induces skin loss, similar to that seen in extensive burns. It is associated with high morbidity and mortality and there is no clear agreement on effective treatment. Objectives To assess the effects of all interventions for the treatment of toxic epidermal necrolysis. Search methods We searched the Cochrane Skin Group Specialised Register (March 2001), the Cochrane Controlled Trials Register (March 2001), MEDLINE (1966 to December 2001), EMBASE (1980 to December 2001), DARE (4th Quarter 2001) and CINAHL (1982 to October 2001). Selection criteria Randomised controlled trials of therapeutic and supportive interventions that included participants clinically diagnosed with toxic epidermal necrolysis were included. Data collection and analysis Two independent authors carried out study selection and assessment of methodological quality. Only one randomised controlled trial of treatment was identified. This trial compared the effectiveness of thalidomide with placebo and included 22 patients, 12 in the treatment group and 10 in the placebo group. Patients on the treatment arm received thalidomide 200 mg twice daily for 5 days. The main end point was the measurement of the progression of skin detachment after seven days. Other end points were the overall mortality and severity of the disease evaluated with the simplified acute physiology score. The study was terminated as the mortality on the treatment arm was 83% compared to 30% on the control arm (relative risk 2.78, 95% confidence interval 1.04 to 7.40). No randomised controlled trials of the most commonly used current treatments i.e. systemic steroids, cyclosporin A and intravenous immunoglobulins were found. Treatment with thalidomide was not shown to be effective and was associated with significantly higher mortality than placebo. There is no reliable evidence on which to base treatment for toxic epidermal necrolysis, a disease commonly associated with mortality rates of around 30%. More research is required to understand the mechanisms of toxic epidermal necrolysis. International multi‐centre studies are needed in the form of randomised controlled trials, to evaluate treatments for toxic epidermal necrolysis, especially those using high doses of steroid and intravenous immunoglobulins.
t138	t138_7	no	Thalidomide is not safe or effective for the skin condition toxic epidermal necrolysis, but there is not enough evidence to show which treatments are effective.	Toxic epidermal necrolysis is a rare condition where a drug reaction induces skin loss, similar to that seen in extensive burns. It is associated with high morbidity and mortality and there is no clear agreement on effective treatment. Objectives To assess the effects of all interventions for the treatment of toxic epidermal necrolysis. Search methods We searched the Cochrane Skin Group Specialised Register (March 2001), the Cochrane Controlled Trials Register (March 2001), MEDLINE (1966 to December 2001), EMBASE (1980 to December 2001), DARE (4th Quarter 2001) and CINAHL (1982 to October 2001). Selection criteria Randomised controlled trials of therapeutic and supportive interventions that included participants clinically diagnosed with toxic epidermal necrolysis were included. Data collection and analysis Two independent authors carried out study selection and assessment of methodological quality. Only one randomised controlled trial of treatment was identified. This trial compared the effectiveness of thalidomide with placebo and included 22 patients, 12 in the treatment group and 10 in the placebo group. Patients on the treatment arm received thalidomide 200 mg twice daily for 5 days. The main end point was the measurement of the progression of skin detachment after seven days. Other end points were the overall mortality and severity of the disease evaluated with the simplified acute physiology score. The study was terminated as the mortality on the treatment arm was 83% compared to 30% on the control arm (relative risk 2.78, 95% confidence interval 1.04 to 7.40). No randomised controlled trials of the most commonly used current treatments i.e. systemic steroids, cyclosporin A and intravenous immunoglobulins were found. Treatment with thalidomide was not shown to be effective and was associated with significantly higher mortality than placebo. There is no reliable evidence on which to base treatment for toxic epidermal necrolysis, a disease commonly associated with mortality rates of around 30%. More research is required to understand the mechanisms of toxic epidermal necrolysis. International multi‐centre studies are needed in the form of randomised controlled trials, to evaluate treatments for toxic epidermal necrolysis, especially those using high doses of steroid and intravenous immunoglobulins.
t139	t139_1	yes	Neovascular age‐related macular degeneration (AMD) is a progressive and chronic disease of the eye, and a leading cause of severe blindness in elderly populations.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_2	yes	The disease is characterised by the abnormal growth of arteries and veins (neovascularisation) in the macula, a region of the retina (back portion of eye) responsible for central vision.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_3	yes	Without treatment, the leakage of these blood vessels causes swelling and damage to the macula, resulting in a fibrous scar that impairs eyesight.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_4	yes	Approximately one out of 10 people with neovascular AMD suffer legal blindness, accounting for 90% of all cases of severe vision loss due to AMD.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_5	no	Therapies against neovascular AMD target new blood vessels.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_6	no	Bevacizumab (commercial name Avastin®) and ranibizumab (Lucentis®) are biological drugs that bind to and block the function of vascular endothelial growth factor (VEGF), a protein released by cells in the body that stimulates the growth and leakage of blood vessels.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_7	no	The two drugs, accordingly, inhibit the process of neovascularisation.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_8	no	Ranibizumab is approved to treat neovascular AMD by injection into the eye (intravitreal injection), while bevacizumab is approved as a cancer therapy by injection into the vein through the skin.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_9	no	The two drugs have similar chemical structures and the same mechanism of action.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_10	yes	Although their benefits are equivalent, it has been hypothesised that the two drugs have different systemic safety profiles, such that one drug might cause more adverse events (harms) at the level of whole body compared to the other.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_11	no	We evaluated whether the two drugs differed in terms of deaths or serious systemic adverse events (SSAEs) in people with neovascular AMD.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_12	yes	The latter refers to medically related events that result in death, are life‐threatening, require hospital admission or prolong hospital stay, or cause persistent or significant disability.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_13	no	We included nine randomised controlled trials (RCTs), none supported by industry, with 3665 participants directly comparing bevacizumab with ranibizumab.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_14	yes	Six RCTs were completed and published, two RCTs were completed, but unpublished, and one was still in progress.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_15	no	We were able to include safety information from all trials, accessing both published and unpublished data.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_16	yes	Drugs were administered for up to two years according to continuous or discontinuous treatment.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_17	yes	In the first, drugs were regularly administered, irrespective of the remission or progression of the disease; the latter involved 'as needed' (pro re nata, PRN) or 'treat‐and‐extend' regimens in which the drug was injected less frequently as long as there was no recurrence of neovascular manifestations.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_18	no	Follow‐up for adverse events occurred at regular intervals up to one or two years, irrespective of continuous or discontinuous treatment.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_19	yes	All studies used the approved dosage of ranibizumab (0.5 mg) according to the 'Summary of Product Characteristics', and the dosage of bevacizumab most recommended by ophthalmologists for intravitreal injection (1.25 mg).	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_20	no	Three studies excluded patients at high cardiovascular risk.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_21	yes	However, four RCTs considered patients at different cardiovascular risks, representing a wide spectrum of risks and routine practice in hospital settings.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_22	no	Our review found the systemic safety of bevacizumab for neovascular AMD to be similar to that of ranibizumab, except for gastrointestinal disorders, which was a part of a secondary analysis.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t139	t139_23	yes	Deaths are likely to be unrelated to the administration of drugs.	Neovascular age‐related macular degeneration (AMD) is the leading cause of legal blindness in elderly populations of industrialised countries. Bevacizumab (Avastin®) and ranibizumab (Lucentis®) are targeted biological drugs (a monoclonal antibody) that inhibit vascular endothelial growth factor, an angiogenic cytokine that promotes vascular leakage and growth, thereby preventing its pathological angiogenesis. Ranibizumab is approved for intravitreal use to treat neovascular AMD, while bevacizumab is approved for intravenous use as a cancer therapy. However, due to the biological similarity of the two drugs, bevacizumab is widely used off‐label to treat neovascular AMD. Objectives To assess the systemic safety of intravitreal bevacizumab (brand name Avastin®; Genentech/Roche) compared with intravitreal ranibizumab (brand name Lucentis®; Novartis/Genentech) in people with neovascular AMD. Primary outcomes were death and All serious systemic adverse events (All SSAEs), the latter as a composite outcome in accordance with the International Conference on Harmonisation Good Clinical Practice. Secondary outcomes examined specific SSAEs: fatal and non‐fatal myocardial infarctions, strokes, arteriothrombotic events, serious infections, and events grouped in some Medical Dictionary for Regulatory Activities System Organ Classes (MedDRA SOC). We assessed the safety at the longest available follow‐up to a maximum of two years. Search methods We searched CENTRAL, MEDLINE, EMBASE and other online databases up to 27 March 2014. We also searched abstracts and clinical study presentations at meetings, trial registries, and contacted authors of included studies when we had questions. Selection criteria Randomised controlled trials (RCTs) directly comparing intravitreal bevacizumab (1.25 mg) and ranibizumab (0.5 mg) in people with neovascular AMD, regardless of publication status, drug dose, treatment regimen, or follow‐up length, and whether the SSAEs of interest were reported in the trial report. Data collection and analysis Two authors independently selected studies and assessed the risk of bias for each study. Three authors independently extracted data. We conducted random‐effects meta‐analyses for the primary and secondary outcomes. We planned a pre‐specified analysis to explore deaths and All SSAEs at the one‐year follow‐up. We included data from nine studies (3665 participants), including six published (2745 participants) and three unpublished (920 participants) RCTs, none supported by industry. Three studies excluded participants at high cardiovascular risk, increasing clinical heterogeneity among studies. The studies were well designed, and we did not downgrade the quality of the evidence for any of the outcomes due to risk of bias. Although the estimated effects of bevacizumab and ranibizumab on our outcomes were similar, we downgraded the quality of the evidence due to imprecision. At the maximum follow‐up (one or two years), the estimated risk ratio (RR) of death with bevacizumab compared with ranibizumab was 1.10 (95% confidence interval (CI) 0.78 to 1.57, P value = 0.59; eight studies, 3338 participants; moderate quality evidence). Based on the event rates in the studies, this gives a risk of death with ranibizumab of 3.4% and with bevacizumab of 3.7% (95% CI 2.7% to 5.3%). For All SSAEs, the estimated RR was 1.08 (95% CI 0.90 to 1.31, P value = 0.41; nine studies, 3665 participants; low quality evidence). Based on the event rates in the studies, this gives a risk of SSAEs of 22.2% with ranibizumab and with bevacizumab of 24% (95% CI 20% to 29.1%). For the secondary outcomes, we could not detect any difference between bevacizumab and ranibizumab, with the exception of gastrointestinal disorders MedDRA SOC where there was a higher risk with bevacizumab (RR 1.82; 95% CI 1.04 to 3.19, P value = 0.04; six studies, 3190 participants). Pre‐specified analyses of deaths and All SSAEs at one‐year follow‐up did not substantially alter the findings of our review. Fixed‐effect analysis for deaths did not substantially alter the findings of our review, but fixed‐effect analysis of All SSAEs showed an increased risk for bevacizumab (RR 1.12; 95% CI 1.00 to 1.26, P value = 0.04; nine studies, 3665 participants): the meta‐analysis was dominated by a single study (weight = 46.9%). The available evidence was sensitive to the exclusion of CATT or unpublished results. For All SSAEs, the exclusion of CATT moved the overall estimate towards no difference (RR 1.01; 95% CI 0.82 to 1.25, P value = 0.92), while the exclusion of LUCAS yielded a larger RR, with more SSAEs in the bevacizumab group, largely driven by CATT (RR 1.19; 95% CI 1.06 to 1.34, P value = 0.004). The exclusion of all unpublished studies produced a RR of 1.12 for death (95% CI 0.78 to 1.62, P value = 0.53) and a RR of 1.21 for SSAEs (95% CI 1.06 to 1.37, P value = 0.004), indicating a higher risk of SSAEs in those assigned to bevacizumab than ranibizumab. This systematic review of non‐industry sponsored RCTs could not determine a difference between intravitreal bevacizumab and ranibizumab for deaths, All SSAEs, or specific subsets of SSAEs in the first two years of treatment, with the exception of gastrointestinal disorders. The current evidence is imprecise and might vary across levels of patient risks, but overall suggests that if a difference exists, it is likely to be small. Health policies for the utilisation of ranibizumab instead of bevacizumab as a routine intervention for neovascular AMD for reasons of systemic safety are not sustained by evidence. The and quality of evidence should be verified once all trials are fully published.
t140	t140_1	no	Seizures (epileptic attacks) after stroke are a major clinical problem.	This is an updated version of the original Cochrane review published in 2010, Issue 1. Seizures after stroke are an important clinical problem, and they may be associated with poor outcome. The effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke remain unclear. Objectives We aimed to assess the effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke. Search methods We searched the Specialised Registers of the Cochrane Epilepsy Group (12 August 2013) and the Cochrane Stroke Group (12 August 2013), the Cochrane Central Register of Controlled Trials (CENTRAL, The Cochrane Library 2013, Issue 7), and MEDLINE (OVID, 1946 to 12 August 2013). We also checked the reference lists of articles retrieved from these searches. Selection criteria Randomised and quasi‐randomised controlled trials in which participants were assigned to treatment or control group (placebo or no drug). Data collection and analysis Two review authors independently screened all the titles, abstracts, and keywords of publications identified by the searches to assess their eligibility, and both review authors assessed their suitability for inclusion according to prespecified selection criteria. We included only one study for data collection and analysis. We found only one trial that fulfilled the study inclusion criteria of comparison of the effects of an antiepileptic drug with placebo (or no drug) for the primary or secondary prevention of seizures after stroke. This was a prospective randomised, double‐blind, placebo‐controlled trial comparing valproic acid with placebo for primary prevention of seizures in 72 adults (over 18 years of age) with spontaneous non‐aneurysmal, non‐traumatic intracerebral haemorrhage; no statistically significant difference in outcome (seizure occurrence at one year) was demonstrated between groups. Currently, there is insufficient evidence to support the routine use of antiepileptic drugs for the primary or secondary prevention of seizures after stroke. Further well‐conducted research is needed for this important clinical problem.
t140	t140_2	no	It is unclear whether antiepileptic drugs are effective in preventing seizures after stroke in adults.	This is an updated version of the original Cochrane review published in 2010, Issue 1. Seizures after stroke are an important clinical problem, and they may be associated with poor outcome. The effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke remain unclear. Objectives We aimed to assess the effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke. Search methods We searched the Specialised Registers of the Cochrane Epilepsy Group (12 August 2013) and the Cochrane Stroke Group (12 August 2013), the Cochrane Central Register of Controlled Trials (CENTRAL, The Cochrane Library 2013, Issue 7), and MEDLINE (OVID, 1946 to 12 August 2013). We also checked the reference lists of articles retrieved from these searches. Selection criteria Randomised and quasi‐randomised controlled trials in which participants were assigned to treatment or control group (placebo or no drug). Data collection and analysis Two review authors independently screened all the titles, abstracts, and keywords of publications identified by the searches to assess their eligibility, and both review authors assessed their suitability for inclusion according to prespecified selection criteria. We included only one study for data collection and analysis. We found only one trial that fulfilled the study inclusion criteria of comparison of the effects of an antiepileptic drug with placebo (or no drug) for the primary or secondary prevention of seizures after stroke. This was a prospective randomised, double‐blind, placebo‐controlled trial comparing valproic acid with placebo for primary prevention of seizures in 72 adults (over 18 years of age) with spontaneous non‐aneurysmal, non‐traumatic intracerebral haemorrhage; no statistically significant difference in outcome (seizure occurrence at one year) was demonstrated between groups. Currently, there is insufficient evidence to support the routine use of antiepileptic drugs for the primary or secondary prevention of seizures after stroke. Further well‐conducted research is needed for this important clinical problem.
t140	t140_3	yes	We found only one high quality clinical trial that looked at whether antiepileptic drugs may be more effective than placebo in preventing seizures after stroke.	This is an updated version of the original Cochrane review published in 2010, Issue 1. Seizures after stroke are an important clinical problem, and they may be associated with poor outcome. The effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke remain unclear. Objectives We aimed to assess the effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke. Search methods We searched the Specialised Registers of the Cochrane Epilepsy Group (12 August 2013) and the Cochrane Stroke Group (12 August 2013), the Cochrane Central Register of Controlled Trials (CENTRAL, The Cochrane Library 2013, Issue 7), and MEDLINE (OVID, 1946 to 12 August 2013). We also checked the reference lists of articles retrieved from these searches. Selection criteria Randomised and quasi‐randomised controlled trials in which participants were assigned to treatment or control group (placebo or no drug). Data collection and analysis Two review authors independently screened all the titles, abstracts, and keywords of publications identified by the searches to assess their eligibility, and both review authors assessed their suitability for inclusion according to prespecified selection criteria. We included only one study for data collection and analysis. We found only one trial that fulfilled the study inclusion criteria of comparison of the effects of an antiepileptic drug with placebo (or no drug) for the primary or secondary prevention of seizures after stroke. This was a prospective randomised, double‐blind, placebo‐controlled trial comparing valproic acid with placebo for primary prevention of seizures in 72 adults (over 18 years of age) with spontaneous non‐aneurysmal, non‐traumatic intracerebral haemorrhage; no statistically significant difference in outcome (seizure occurrence at one year) was demonstrated between groups. Currently, there is insufficient evidence to support the routine use of antiepileptic drugs for the primary or secondary prevention of seizures after stroke. Further well‐conducted research is needed for this important clinical problem.
t140	t140_4	no	This was a prospective randomised, double‐blind, placebo controlled trial studying the efficacy of valproic acid versus placebo in the primary prevention of seizure in 72 adults (over 18 years of age) with spontaneous non‐aneurysmal, non‐traumatic intracerebral haemorrhage.	This is an updated version of the original Cochrane review published in 2010, Issue 1. Seizures after stroke are an important clinical problem, and they may be associated with poor outcome. The effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke remain unclear. Objectives We aimed to assess the effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke. Search methods We searched the Specialised Registers of the Cochrane Epilepsy Group (12 August 2013) and the Cochrane Stroke Group (12 August 2013), the Cochrane Central Register of Controlled Trials (CENTRAL, The Cochrane Library 2013, Issue 7), and MEDLINE (OVID, 1946 to 12 August 2013). We also checked the reference lists of articles retrieved from these searches. Selection criteria Randomised and quasi‐randomised controlled trials in which participants were assigned to treatment or control group (placebo or no drug). Data collection and analysis Two review authors independently screened all the titles, abstracts, and keywords of publications identified by the searches to assess their eligibility, and both review authors assessed their suitability for inclusion according to prespecified selection criteria. We included only one study for data collection and analysis. We found only one trial that fulfilled the study inclusion criteria of comparison of the effects of an antiepileptic drug with placebo (or no drug) for the primary or secondary prevention of seizures after stroke. This was a prospective randomised, double‐blind, placebo‐controlled trial comparing valproic acid with placebo for primary prevention of seizures in 72 adults (over 18 years of age) with spontaneous non‐aneurysmal, non‐traumatic intracerebral haemorrhage; no statistically significant difference in outcome (seizure occurrence at one year) was demonstrated between groups. Currently, there is insufficient evidence to support the routine use of antiepileptic drugs for the primary or secondary prevention of seizures after stroke. Further well‐conducted research is needed for this important clinical problem.
t140	t140_5	yes	Patients were randomly allocated to either the treatment or the placebo group with active treatment lasting one month; the primary outcome was seizure occurrence at one year.	This is an updated version of the original Cochrane review published in 2010, Issue 1. Seizures after stroke are an important clinical problem, and they may be associated with poor outcome. The effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke remain unclear. Objectives We aimed to assess the effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke. Search methods We searched the Specialised Registers of the Cochrane Epilepsy Group (12 August 2013) and the Cochrane Stroke Group (12 August 2013), the Cochrane Central Register of Controlled Trials (CENTRAL, The Cochrane Library 2013, Issue 7), and MEDLINE (OVID, 1946 to 12 August 2013). We also checked the reference lists of articles retrieved from these searches. Selection criteria Randomised and quasi‐randomised controlled trials in which participants were assigned to treatment or control group (placebo or no drug). Data collection and analysis Two review authors independently screened all the titles, abstracts, and keywords of publications identified by the searches to assess their eligibility, and both review authors assessed their suitability for inclusion according to prespecified selection criteria. We included only one study for data collection and analysis. We found only one trial that fulfilled the study inclusion criteria of comparison of the effects of an antiepileptic drug with placebo (or no drug) for the primary or secondary prevention of seizures after stroke. This was a prospective randomised, double‐blind, placebo‐controlled trial comparing valproic acid with placebo for primary prevention of seizures in 72 adults (over 18 years of age) with spontaneous non‐aneurysmal, non‐traumatic intracerebral haemorrhage; no statistically significant difference in outcome (seizure occurrence at one year) was demonstrated between groups. Currently, there is insufficient evidence to support the routine use of antiepileptic drugs for the primary or secondary prevention of seizures after stroke. Further well‐conducted research is needed for this important clinical problem.
t140	t140_6	yes	People with very early seizures (within 24 hours of onset of haemorrhage) were excluded from the study.	This is an updated version of the original Cochrane review published in 2010, Issue 1. Seizures after stroke are an important clinical problem, and they may be associated with poor outcome. The effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke remain unclear. Objectives We aimed to assess the effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke. Search methods We searched the Specialised Registers of the Cochrane Epilepsy Group (12 August 2013) and the Cochrane Stroke Group (12 August 2013), the Cochrane Central Register of Controlled Trials (CENTRAL, The Cochrane Library 2013, Issue 7), and MEDLINE (OVID, 1946 to 12 August 2013). We also checked the reference lists of articles retrieved from these searches. Selection criteria Randomised and quasi‐randomised controlled trials in which participants were assigned to treatment or control group (placebo or no drug). Data collection and analysis Two review authors independently screened all the titles, abstracts, and keywords of publications identified by the searches to assess their eligibility, and both review authors assessed their suitability for inclusion according to prespecified selection criteria. We included only one study for data collection and analysis. We found only one trial that fulfilled the study inclusion criteria of comparison of the effects of an antiepileptic drug with placebo (or no drug) for the primary or secondary prevention of seizures after stroke. This was a prospective randomised, double‐blind, placebo‐controlled trial comparing valproic acid with placebo for primary prevention of seizures in 72 adults (over 18 years of age) with spontaneous non‐aneurysmal, non‐traumatic intracerebral haemorrhage; no statistically significant difference in outcome (seizure occurrence at one year) was demonstrated between groups. Currently, there is insufficient evidence to support the routine use of antiepileptic drugs for the primary or secondary prevention of seizures after stroke. Further well‐conducted research is needed for this important clinical problem.
t140	t140_7	yes	Seizure was diagnosed on the basis of eye‐witness evidence from staff, relatives or other eye witnesses.	This is an updated version of the original Cochrane review published in 2010, Issue 1. Seizures after stroke are an important clinical problem, and they may be associated with poor outcome. The effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke remain unclear. Objectives We aimed to assess the effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke. Search methods We searched the Specialised Registers of the Cochrane Epilepsy Group (12 August 2013) and the Cochrane Stroke Group (12 August 2013), the Cochrane Central Register of Controlled Trials (CENTRAL, The Cochrane Library 2013, Issue 7), and MEDLINE (OVID, 1946 to 12 August 2013). We also checked the reference lists of articles retrieved from these searches. Selection criteria Randomised and quasi‐randomised controlled trials in which participants were assigned to treatment or control group (placebo or no drug). Data collection and analysis Two review authors independently screened all the titles, abstracts, and keywords of publications identified by the searches to assess their eligibility, and both review authors assessed their suitability for inclusion according to prespecified selection criteria. We included only one study for data collection and analysis. We found only one trial that fulfilled the study inclusion criteria of comparison of the effects of an antiepileptic drug with placebo (or no drug) for the primary or secondary prevention of seizures after stroke. This was a prospective randomised, double‐blind, placebo‐controlled trial comparing valproic acid with placebo for primary prevention of seizures in 72 adults (over 18 years of age) with spontaneous non‐aneurysmal, non‐traumatic intracerebral haemorrhage; no statistically significant difference in outcome (seizure occurrence at one year) was demonstrated between groups. Currently, there is insufficient evidence to support the routine use of antiepileptic drugs for the primary or secondary prevention of seizures after stroke. Further well‐conducted research is needed for this important clinical problem.
t140	t140_8	yes	There does not appear to be bias in Gilad 2011 , on the basis of the information available within the study.	This is an updated version of the original Cochrane review published in 2010, Issue 1. Seizures after stroke are an important clinical problem, and they may be associated with poor outcome. The effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke remain unclear. Objectives We aimed to assess the effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke. Search methods We searched the Specialised Registers of the Cochrane Epilepsy Group (12 August 2013) and the Cochrane Stroke Group (12 August 2013), the Cochrane Central Register of Controlled Trials (CENTRAL, The Cochrane Library 2013, Issue 7), and MEDLINE (OVID, 1946 to 12 August 2013). We also checked the reference lists of articles retrieved from these searches. Selection criteria Randomised and quasi‐randomised controlled trials in which participants were assigned to treatment or control group (placebo or no drug). Data collection and analysis Two review authors independently screened all the titles, abstracts, and keywords of publications identified by the searches to assess their eligibility, and both review authors assessed their suitability for inclusion according to prespecified selection criteria. We included only one study for data collection and analysis. We found only one trial that fulfilled the study inclusion criteria of comparison of the effects of an antiepileptic drug with placebo (or no drug) for the primary or secondary prevention of seizures after stroke. This was a prospective randomised, double‐blind, placebo‐controlled trial comparing valproic acid with placebo for primary prevention of seizures in 72 adults (over 18 years of age) with spontaneous non‐aneurysmal, non‐traumatic intracerebral haemorrhage; no statistically significant difference in outcome (seizure occurrence at one year) was demonstrated between groups. Currently, there is insufficient evidence to support the routine use of antiepileptic drugs for the primary or secondary prevention of seizures after stroke. Further well‐conducted research is needed for this important clinical problem.
t140	t140_9	yes	Gilad 2011 did not show a statistically significant benefit when comparing valproic acid with placebo for the primary prevention of seizures after spontaneous non‐aneurysmal, non‐traumatic intracerebral haemorrhage.	This is an updated version of the original Cochrane review published in 2010, Issue 1. Seizures after stroke are an important clinical problem, and they may be associated with poor outcome. The effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke remain unclear. Objectives We aimed to assess the effects of antiepileptic drugs for the primary and secondary prevention of seizures after stroke. Search methods We searched the Specialised Registers of the Cochrane Epilepsy Group (12 August 2013) and the Cochrane Stroke Group (12 August 2013), the Cochrane Central Register of Controlled Trials (CENTRAL, The Cochrane Library 2013, Issue 7), and MEDLINE (OVID, 1946 to 12 August 2013). We also checked the reference lists of articles retrieved from these searches. Selection criteria Randomised and quasi‐randomised controlled trials in which participants were assigned to treatment or control group (placebo or no drug). Data collection and analysis Two review authors independently screened all the titles, abstracts, and keywords of publications identified by the searches to assess their eligibility, and both review authors assessed their suitability for inclusion according to prespecified selection criteria. We included only one study for data collection and analysis. We found only one trial that fulfilled the study inclusion criteria of comparison of the effects of an antiepileptic drug with placebo (or no drug) for the primary or secondary prevention of seizures after stroke. This was a prospective randomised, double‐blind, placebo‐controlled trial comparing valproic acid with placebo for primary prevention of seizures in 72 adults (over 18 years of age) with spontaneous non‐aneurysmal, non‐traumatic intracerebral haemorrhage; no statistically significant difference in outcome (seizure occurrence at one year) was demonstrated between groups. Currently, there is insufficient evidence to support the routine use of antiepileptic drugs for the primary or secondary prevention of seizures after stroke. Further well‐conducted research is needed for this important clinical problem.
t141	t141_1	yes	As many as half of all children with tic disorders (a combination of repetitive motions vocalizations), also have ADHD (issues with hyperactivity, impulsivity and maintaining attention).	This is an update of the original Cochrane Review published in Issue 4, 2011. Attention deficit hyperactivity disorder (ADHD) is the most prevalent of the comorbid psychiatric disorders that complicate tic disorders. Medications commonly used to treat ADHD symptoms include stimulants such as methylphenidate and amphetamine; non‐stimulants, such as atomoxetine; tricyclic antidepressants; and alpha agonists. Alpha agonists are also used as a treatment for tics. Due to the impact of ADHD symptoms on the child with tic disorder, treatment of ADHD is often of greater priority than the medical management of tics. However, for many decades, clinicians have been reluctant to use stimulants to treat children with ADHD and tics for fear of worsening their tics. Objectives To assess the effects of pharmacological treatments for ADHD in children with comorbid tic disorders on symptoms of ADHD and tics. Search methods In September 2017, we searched CENTRAL, MEDLINE, Embase, and 12 other databases. We also searched two trial registers and contacted experts in the field for any ongoing or unpublished studies. Selection criteria We included randomized, double‐blind, controlled trials of any pharmacological treatment for ADHD used specifically in children with comorbid tic disorders. We included both parallel‐group and cross‐over study designs. Data collection and analysis We used standard methodological procedures of Cochrane, in that two review authors independently selected studies, extracted data using standardized forms, assessed risk of bias, and graded the overall quality of the evidence by using the GRADE approach. We included eight randomized controlled trials (four of which were cross‐over trials) with 510 participants (443 boys, 67 girls) in this review. Participants in these studies were children with both ADHD and a chronic tic disorder. All studies took place in the USA and ranged from three to 22 weeks in duration. Five of the eight studies were funded by charitable organizations or government agencies, or both. One study was funded by the drug manufacturer. The other two studies did not specify the source of funding. Risk of bias of included studies was low for blinding; low or unclear for random sequence generation, allocation concealment, and attrition bias; and low or high for selective outcome reporting. We were unable to combine any of the studies in a meta‐analysis due to important clinical heterogeneity and unit‐of‐analysis issues. Several of the trials assessed multiple agents. Medications assessed included methylphenidate, clonidine, desipramine, dextroamphetamine, guanfacine, atomoxetine, and deprenyl. There was low‐quality evidence for methylphenidate, atomoxetine, and clonidine, and very low‐quality evidence for desipramine, dextroamphetamine, guanfacine and deprenyl in the treatment of ADHD in children with tics. All studies, with the exception of a study using deprenyl, reported improvement in symptoms of ADHD. Tic symptoms also improved in children treated with guanfacine, desipramine, methylphenidate, clonidine, and a combination of methylphenidate and clonidine. In one study, tics limited further dosage increases of methylphenidate. High‐dose dextroamphetamine appeared to worsen tics in one study, although the length of this study was limited to three weeks. There was appetite suppression or weight loss in association with methylphenidate, dextroamphetamine, atomoxetine, and desipramine. There was insomnia associated with methylphenidate and dextroamphetamine, and sedation associated with clonidine. Following an updated search of potentially relevant studies, we found no new studies that matched our inclusion criteria and thus our conclusions have not changed. Methylphenidate, clonidine, guanfacine, desipramine, and atomoxetine appear to reduce ADHD symptoms in children with tics though the quality of the available evidence was low to very low. Although stimulants have not been shown to worsen tics in most people with tic disorders, they may, nonetheless, exacerbate tics in individual cases. In these instances, treatment with alpha agonists or atomoxetine may be an alternative. Although there is evidence that desipramine may improve tics and ADHD in children, safety concerns will likely continue to limit its use in this population.
t141	t141_2	no	Symptoms of ADHD are often more disabling for children than their tics.	This is an update of the original Cochrane Review published in Issue 4, 2011. Attention deficit hyperactivity disorder (ADHD) is the most prevalent of the comorbid psychiatric disorders that complicate tic disorders. Medications commonly used to treat ADHD symptoms include stimulants such as methylphenidate and amphetamine; non‐stimulants, such as atomoxetine; tricyclic antidepressants; and alpha agonists. Alpha agonists are also used as a treatment for tics. Due to the impact of ADHD symptoms on the child with tic disorder, treatment of ADHD is often of greater priority than the medical management of tics. However, for many decades, clinicians have been reluctant to use stimulants to treat children with ADHD and tics for fear of worsening their tics. Objectives To assess the effects of pharmacological treatments for ADHD in children with comorbid tic disorders on symptoms of ADHD and tics. Search methods In September 2017, we searched CENTRAL, MEDLINE, Embase, and 12 other databases. We also searched two trial registers and contacted experts in the field for any ongoing or unpublished studies. Selection criteria We included randomized, double‐blind, controlled trials of any pharmacological treatment for ADHD used specifically in children with comorbid tic disorders. We included both parallel‐group and cross‐over study designs. Data collection and analysis We used standard methodological procedures of Cochrane, in that two review authors independently selected studies, extracted data using standardized forms, assessed risk of bias, and graded the overall quality of the evidence by using the GRADE approach. We included eight randomized controlled trials (four of which were cross‐over trials) with 510 participants (443 boys, 67 girls) in this review. Participants in these studies were children with both ADHD and a chronic tic disorder. All studies took place in the USA and ranged from three to 22 weeks in duration. Five of the eight studies were funded by charitable organizations or government agencies, or both. One study was funded by the drug manufacturer. The other two studies did not specify the source of funding. Risk of bias of included studies was low for blinding; low or unclear for random sequence generation, allocation concealment, and attrition bias; and low or high for selective outcome reporting. We were unable to combine any of the studies in a meta‐analysis due to important clinical heterogeneity and unit‐of‐analysis issues. Several of the trials assessed multiple agents. Medications assessed included methylphenidate, clonidine, desipramine, dextroamphetamine, guanfacine, atomoxetine, and deprenyl. There was low‐quality evidence for methylphenidate, atomoxetine, and clonidine, and very low‐quality evidence for desipramine, dextroamphetamine, guanfacine and deprenyl in the treatment of ADHD in children with tics. All studies, with the exception of a study using deprenyl, reported improvement in symptoms of ADHD. Tic symptoms also improved in children treated with guanfacine, desipramine, methylphenidate, clonidine, and a combination of methylphenidate and clonidine. In one study, tics limited further dosage increases of methylphenidate. High‐dose dextroamphetamine appeared to worsen tics in one study, although the length of this study was limited to three weeks. There was appetite suppression or weight loss in association with methylphenidate, dextroamphetamine, atomoxetine, and desipramine. There was insomnia associated with methylphenidate and dextroamphetamine, and sedation associated with clonidine. Following an updated search of potentially relevant studies, we found no new studies that matched our inclusion criteria and thus our conclusions have not changed. Methylphenidate, clonidine, guanfacine, desipramine, and atomoxetine appear to reduce ADHD symptoms in children with tics though the quality of the available evidence was low to very low. Although stimulants have not been shown to worsen tics in most people with tic disorders, they may, nonetheless, exacerbate tics in individual cases. In these instances, treatment with alpha agonists or atomoxetine may be an alternative. Although there is evidence that desipramine may improve tics and ADHD in children, safety concerns will likely continue to limit its use in this population.
t141	t141_3	yes	Historically, the reported ability of stimulant medications to worsen tics has limited their use in children who have both a chronic tic disorder (lasting over a year since the first tic onset) and ADHD.	This is an update of the original Cochrane Review published in Issue 4, 2011. Attention deficit hyperactivity disorder (ADHD) is the most prevalent of the comorbid psychiatric disorders that complicate tic disorders. Medications commonly used to treat ADHD symptoms include stimulants such as methylphenidate and amphetamine; non‐stimulants, such as atomoxetine; tricyclic antidepressants; and alpha agonists. Alpha agonists are also used as a treatment for tics. Due to the impact of ADHD symptoms on the child with tic disorder, treatment of ADHD is often of greater priority than the medical management of tics. However, for many decades, clinicians have been reluctant to use stimulants to treat children with ADHD and tics for fear of worsening their tics. Objectives To assess the effects of pharmacological treatments for ADHD in children with comorbid tic disorders on symptoms of ADHD and tics. Search methods In September 2017, we searched CENTRAL, MEDLINE, Embase, and 12 other databases. We also searched two trial registers and contacted experts in the field for any ongoing or unpublished studies. Selection criteria We included randomized, double‐blind, controlled trials of any pharmacological treatment for ADHD used specifically in children with comorbid tic disorders. We included both parallel‐group and cross‐over study designs. Data collection and analysis We used standard methodological procedures of Cochrane, in that two review authors independently selected studies, extracted data using standardized forms, assessed risk of bias, and graded the overall quality of the evidence by using the GRADE approach. We included eight randomized controlled trials (four of which were cross‐over trials) with 510 participants (443 boys, 67 girls) in this review. Participants in these studies were children with both ADHD and a chronic tic disorder. All studies took place in the USA and ranged from three to 22 weeks in duration. Five of the eight studies were funded by charitable organizations or government agencies, or both. One study was funded by the drug manufacturer. The other two studies did not specify the source of funding. Risk of bias of included studies was low for blinding; low or unclear for random sequence generation, allocation concealment, and attrition bias; and low or high for selective outcome reporting. We were unable to combine any of the studies in a meta‐analysis due to important clinical heterogeneity and unit‐of‐analysis issues. Several of the trials assessed multiple agents. Medications assessed included methylphenidate, clonidine, desipramine, dextroamphetamine, guanfacine, atomoxetine, and deprenyl. There was low‐quality evidence for methylphenidate, atomoxetine, and clonidine, and very low‐quality evidence for desipramine, dextroamphetamine, guanfacine and deprenyl in the treatment of ADHD in children with tics. All studies, with the exception of a study using deprenyl, reported improvement in symptoms of ADHD. Tic symptoms also improved in children treated with guanfacine, desipramine, methylphenidate, clonidine, and a combination of methylphenidate and clonidine. In one study, tics limited further dosage increases of methylphenidate. High‐dose dextroamphetamine appeared to worsen tics in one study, although the length of this study was limited to three weeks. There was appetite suppression or weight loss in association with methylphenidate, dextroamphetamine, atomoxetine, and desipramine. There was insomnia associated with methylphenidate and dextroamphetamine, and sedation associated with clonidine. Following an updated search of potentially relevant studies, we found no new studies that matched our inclusion criteria and thus our conclusions have not changed. Methylphenidate, clonidine, guanfacine, desipramine, and atomoxetine appear to reduce ADHD symptoms in children with tics though the quality of the available evidence was low to very low. Although stimulants have not been shown to worsen tics in most people with tic disorders, they may, nonetheless, exacerbate tics in individual cases. In these instances, treatment with alpha agonists or atomoxetine may be an alternative. Although there is evidence that desipramine may improve tics and ADHD in children, safety concerns will likely continue to limit its use in this population.
t141	t141_4	no	To evaluate evidence for this reported phenomenon, we searched for clinical trials of medications for ADHD used specifically in children with tic disorders.	This is an update of the original Cochrane Review published in Issue 4, 2011. Attention deficit hyperactivity disorder (ADHD) is the most prevalent of the comorbid psychiatric disorders that complicate tic disorders. Medications commonly used to treat ADHD symptoms include stimulants such as methylphenidate and amphetamine; non‐stimulants, such as atomoxetine; tricyclic antidepressants; and alpha agonists. Alpha agonists are also used as a treatment for tics. Due to the impact of ADHD symptoms on the child with tic disorder, treatment of ADHD is often of greater priority than the medical management of tics. However, for many decades, clinicians have been reluctant to use stimulants to treat children with ADHD and tics for fear of worsening their tics. Objectives To assess the effects of pharmacological treatments for ADHD in children with comorbid tic disorders on symptoms of ADHD and tics. Search methods In September 2017, we searched CENTRAL, MEDLINE, Embase, and 12 other databases. We also searched two trial registers and contacted experts in the field for any ongoing or unpublished studies. Selection criteria We included randomized, double‐blind, controlled trials of any pharmacological treatment for ADHD used specifically in children with comorbid tic disorders. We included both parallel‐group and cross‐over study designs. Data collection and analysis We used standard methodological procedures of Cochrane, in that two review authors independently selected studies, extracted data using standardized forms, assessed risk of bias, and graded the overall quality of the evidence by using the GRADE approach. We included eight randomized controlled trials (four of which were cross‐over trials) with 510 participants (443 boys, 67 girls) in this review. Participants in these studies were children with both ADHD and a chronic tic disorder. All studies took place in the USA and ranged from three to 22 weeks in duration. Five of the eight studies were funded by charitable organizations or government agencies, or both. One study was funded by the drug manufacturer. The other two studies did not specify the source of funding. Risk of bias of included studies was low for blinding; low or unclear for random sequence generation, allocation concealment, and attrition bias; and low or high for selective outcome reporting. We were unable to combine any of the studies in a meta‐analysis due to important clinical heterogeneity and unit‐of‐analysis issues. Several of the trials assessed multiple agents. Medications assessed included methylphenidate, clonidine, desipramine, dextroamphetamine, guanfacine, atomoxetine, and deprenyl. There was low‐quality evidence for methylphenidate, atomoxetine, and clonidine, and very low‐quality evidence for desipramine, dextroamphetamine, guanfacine and deprenyl in the treatment of ADHD in children with tics. All studies, with the exception of a study using deprenyl, reported improvement in symptoms of ADHD. Tic symptoms also improved in children treated with guanfacine, desipramine, methylphenidate, clonidine, and a combination of methylphenidate and clonidine. In one study, tics limited further dosage increases of methylphenidate. High‐dose dextroamphetamine appeared to worsen tics in one study, although the length of this study was limited to three weeks. There was appetite suppression or weight loss in association with methylphenidate, dextroamphetamine, atomoxetine, and desipramine. There was insomnia associated with methylphenidate and dextroamphetamine, and sedation associated with clonidine. Following an updated search of potentially relevant studies, we found no new studies that matched our inclusion criteria and thus our conclusions have not changed. Methylphenidate, clonidine, guanfacine, desipramine, and atomoxetine appear to reduce ADHD symptoms in children with tics though the quality of the available evidence was low to very low. Although stimulants have not been shown to worsen tics in most people with tic disorders, they may, nonetheless, exacerbate tics in individual cases. In these instances, treatment with alpha agonists or atomoxetine may be an alternative. Although there is evidence that desipramine may improve tics and ADHD in children, safety concerns will likely continue to limit its use in this population.
t141	t141_5	no	Participants in these studies were children with both ADHD and a chronic tic disorder.	This is an update of the original Cochrane Review published in Issue 4, 2011. Attention deficit hyperactivity disorder (ADHD) is the most prevalent of the comorbid psychiatric disorders that complicate tic disorders. Medications commonly used to treat ADHD symptoms include stimulants such as methylphenidate and amphetamine; non‐stimulants, such as atomoxetine; tricyclic antidepressants; and alpha agonists. Alpha agonists are also used as a treatment for tics. Due to the impact of ADHD symptoms on the child with tic disorder, treatment of ADHD is often of greater priority than the medical management of tics. However, for many decades, clinicians have been reluctant to use stimulants to treat children with ADHD and tics for fear of worsening their tics. Objectives To assess the effects of pharmacological treatments for ADHD in children with comorbid tic disorders on symptoms of ADHD and tics. Search methods In September 2017, we searched CENTRAL, MEDLINE, Embase, and 12 other databases. We also searched two trial registers and contacted experts in the field for any ongoing or unpublished studies. Selection criteria We included randomized, double‐blind, controlled trials of any pharmacological treatment for ADHD used specifically in children with comorbid tic disorders. We included both parallel‐group and cross‐over study designs. Data collection and analysis We used standard methodological procedures of Cochrane, in that two review authors independently selected studies, extracted data using standardized forms, assessed risk of bias, and graded the overall quality of the evidence by using the GRADE approach. We included eight randomized controlled trials (four of which were cross‐over trials) with 510 participants (443 boys, 67 girls) in this review. Participants in these studies were children with both ADHD and a chronic tic disorder. All studies took place in the USA and ranged from three to 22 weeks in duration. Five of the eight studies were funded by charitable organizations or government agencies, or both. One study was funded by the drug manufacturer. The other two studies did not specify the source of funding. Risk of bias of included studies was low for blinding; low or unclear for random sequence generation, allocation concealment, and attrition bias; and low or high for selective outcome reporting. We were unable to combine any of the studies in a meta‐analysis due to important clinical heterogeneity and unit‐of‐analysis issues. Several of the trials assessed multiple agents. Medications assessed included methylphenidate, clonidine, desipramine, dextroamphetamine, guanfacine, atomoxetine, and deprenyl. There was low‐quality evidence for methylphenidate, atomoxetine, and clonidine, and very low‐quality evidence for desipramine, dextroamphetamine, guanfacine and deprenyl in the treatment of ADHD in children with tics. All studies, with the exception of a study using deprenyl, reported improvement in symptoms of ADHD. Tic symptoms also improved in children treated with guanfacine, desipramine, methylphenidate, clonidine, and a combination of methylphenidate and clonidine. In one study, tics limited further dosage increases of methylphenidate. High‐dose dextroamphetamine appeared to worsen tics in one study, although the length of this study was limited to three weeks. There was appetite suppression or weight loss in association with methylphenidate, dextroamphetamine, atomoxetine, and desipramine. There was insomnia associated with methylphenidate and dextroamphetamine, and sedation associated with clonidine. Following an updated search of potentially relevant studies, we found no new studies that matched our inclusion criteria and thus our conclusions have not changed. Methylphenidate, clonidine, guanfacine, desipramine, and atomoxetine appear to reduce ADHD symptoms in children with tics though the quality of the available evidence was low to very low. Although stimulants have not been shown to worsen tics in most people with tic disorders, they may, nonetheless, exacerbate tics in individual cases. In these instances, treatment with alpha agonists or atomoxetine may be an alternative. Although there is evidence that desipramine may improve tics and ADHD in children, safety concerns will likely continue to limit its use in this population.
t141	t141_6	no	The included studies evaluated several different medications for ADHD, including stimulants (methylphenidate, dextroamphetamine) and non‐stimulants (clonidine, guanfacine, desipramine, atomoxetine, and deprenyl).	This is an update of the original Cochrane Review published in Issue 4, 2011. Attention deficit hyperactivity disorder (ADHD) is the most prevalent of the comorbid psychiatric disorders that complicate tic disorders. Medications commonly used to treat ADHD symptoms include stimulants such as methylphenidate and amphetamine; non‐stimulants, such as atomoxetine; tricyclic antidepressants; and alpha agonists. Alpha agonists are also used as a treatment for tics. Due to the impact of ADHD symptoms on the child with tic disorder, treatment of ADHD is often of greater priority than the medical management of tics. However, for many decades, clinicians have been reluctant to use stimulants to treat children with ADHD and tics for fear of worsening their tics. Objectives To assess the effects of pharmacological treatments for ADHD in children with comorbid tic disorders on symptoms of ADHD and tics. Search methods In September 2017, we searched CENTRAL, MEDLINE, Embase, and 12 other databases. We also searched two trial registers and contacted experts in the field for any ongoing or unpublished studies. Selection criteria We included randomized, double‐blind, controlled trials of any pharmacological treatment for ADHD used specifically in children with comorbid tic disorders. We included both parallel‐group and cross‐over study designs. Data collection and analysis We used standard methodological procedures of Cochrane, in that two review authors independently selected studies, extracted data using standardized forms, assessed risk of bias, and graded the overall quality of the evidence by using the GRADE approach. We included eight randomized controlled trials (four of which were cross‐over trials) with 510 participants (443 boys, 67 girls) in this review. Participants in these studies were children with both ADHD and a chronic tic disorder. All studies took place in the USA and ranged from three to 22 weeks in duration. Five of the eight studies were funded by charitable organizations or government agencies, or both. One study was funded by the drug manufacturer. The other two studies did not specify the source of funding. Risk of bias of included studies was low for blinding; low or unclear for random sequence generation, allocation concealment, and attrition bias; and low or high for selective outcome reporting. We were unable to combine any of the studies in a meta‐analysis due to important clinical heterogeneity and unit‐of‐analysis issues. Several of the trials assessed multiple agents. Medications assessed included methylphenidate, clonidine, desipramine, dextroamphetamine, guanfacine, atomoxetine, and deprenyl. There was low‐quality evidence for methylphenidate, atomoxetine, and clonidine, and very low‐quality evidence for desipramine, dextroamphetamine, guanfacine and deprenyl in the treatment of ADHD in children with tics. All studies, with the exception of a study using deprenyl, reported improvement in symptoms of ADHD. Tic symptoms also improved in children treated with guanfacine, desipramine, methylphenidate, clonidine, and a combination of methylphenidate and clonidine. In one study, tics limited further dosage increases of methylphenidate. High‐dose dextroamphetamine appeared to worsen tics in one study, although the length of this study was limited to three weeks. There was appetite suppression or weight loss in association with methylphenidate, dextroamphetamine, atomoxetine, and desipramine. There was insomnia associated with methylphenidate and dextroamphetamine, and sedation associated with clonidine. Following an updated search of potentially relevant studies, we found no new studies that matched our inclusion criteria and thus our conclusions have not changed. Methylphenidate, clonidine, guanfacine, desipramine, and atomoxetine appear to reduce ADHD symptoms in children with tics though the quality of the available evidence was low to very low. Although stimulants have not been shown to worsen tics in most people with tic disorders, they may, nonetheless, exacerbate tics in individual cases. In these instances, treatment with alpha agonists or atomoxetine may be an alternative. Although there is evidence that desipramine may improve tics and ADHD in children, safety concerns will likely continue to limit its use in this population.
t141	t141_7	no	Study funding sources Five of the eight studies were funded by charitable organizations or government agencies, or both.	This is an update of the original Cochrane Review published in Issue 4, 2011. Attention deficit hyperactivity disorder (ADHD) is the most prevalent of the comorbid psychiatric disorders that complicate tic disorders. Medications commonly used to treat ADHD symptoms include stimulants such as methylphenidate and amphetamine; non‐stimulants, such as atomoxetine; tricyclic antidepressants; and alpha agonists. Alpha agonists are also used as a treatment for tics. Due to the impact of ADHD symptoms on the child with tic disorder, treatment of ADHD is often of greater priority than the medical management of tics. However, for many decades, clinicians have been reluctant to use stimulants to treat children with ADHD and tics for fear of worsening their tics. Objectives To assess the effects of pharmacological treatments for ADHD in children with comorbid tic disorders on symptoms of ADHD and tics. Search methods In September 2017, we searched CENTRAL, MEDLINE, Embase, and 12 other databases. We also searched two trial registers and contacted experts in the field for any ongoing or unpublished studies. Selection criteria We included randomized, double‐blind, controlled trials of any pharmacological treatment for ADHD used specifically in children with comorbid tic disorders. We included both parallel‐group and cross‐over study designs. Data collection and analysis We used standard methodological procedures of Cochrane, in that two review authors independently selected studies, extracted data using standardized forms, assessed risk of bias, and graded the overall quality of the evidence by using the GRADE approach. We included eight randomized controlled trials (four of which were cross‐over trials) with 510 participants (443 boys, 67 girls) in this review. Participants in these studies were children with both ADHD and a chronic tic disorder. All studies took place in the USA and ranged from three to 22 weeks in duration. Five of the eight studies were funded by charitable organizations or government agencies, or both. One study was funded by the drug manufacturer. The other two studies did not specify the source of funding. Risk of bias of included studies was low for blinding; low or unclear for random sequence generation, allocation concealment, and attrition bias; and low or high for selective outcome reporting. We were unable to combine any of the studies in a meta‐analysis due to important clinical heterogeneity and unit‐of‐analysis issues. Several of the trials assessed multiple agents. Medications assessed included methylphenidate, clonidine, desipramine, dextroamphetamine, guanfacine, atomoxetine, and deprenyl. There was low‐quality evidence for methylphenidate, atomoxetine, and clonidine, and very low‐quality evidence for desipramine, dextroamphetamine, guanfacine and deprenyl in the treatment of ADHD in children with tics. All studies, with the exception of a study using deprenyl, reported improvement in symptoms of ADHD. Tic symptoms also improved in children treated with guanfacine, desipramine, methylphenidate, clonidine, and a combination of methylphenidate and clonidine. In one study, tics limited further dosage increases of methylphenidate. High‐dose dextroamphetamine appeared to worsen tics in one study, although the length of this study was limited to three weeks. There was appetite suppression or weight loss in association with methylphenidate, dextroamphetamine, atomoxetine, and desipramine. There was insomnia associated with methylphenidate and dextroamphetamine, and sedation associated with clonidine. Following an updated search of potentially relevant studies, we found no new studies that matched our inclusion criteria and thus our conclusions have not changed. Methylphenidate, clonidine, guanfacine, desipramine, and atomoxetine appear to reduce ADHD symptoms in children with tics though the quality of the available evidence was low to very low. Although stimulants have not been shown to worsen tics in most people with tic disorders, they may, nonetheless, exacerbate tics in individual cases. In these instances, treatment with alpha agonists or atomoxetine may be an alternative. Although there is evidence that desipramine may improve tics and ADHD in children, safety concerns will likely continue to limit its use in this population.
t141	t141_8	no	One study was funded by the drug manufacturer.	This is an update of the original Cochrane Review published in Issue 4, 2011. Attention deficit hyperactivity disorder (ADHD) is the most prevalent of the comorbid psychiatric disorders that complicate tic disorders. Medications commonly used to treat ADHD symptoms include stimulants such as methylphenidate and amphetamine; non‐stimulants, such as atomoxetine; tricyclic antidepressants; and alpha agonists. Alpha agonists are also used as a treatment for tics. Due to the impact of ADHD symptoms on the child with tic disorder, treatment of ADHD is often of greater priority than the medical management of tics. However, for many decades, clinicians have been reluctant to use stimulants to treat children with ADHD and tics for fear of worsening their tics. Objectives To assess the effects of pharmacological treatments for ADHD in children with comorbid tic disorders on symptoms of ADHD and tics. Search methods In September 2017, we searched CENTRAL, MEDLINE, Embase, and 12 other databases. We also searched two trial registers and contacted experts in the field for any ongoing or unpublished studies. Selection criteria We included randomized, double‐blind, controlled trials of any pharmacological treatment for ADHD used specifically in children with comorbid tic disorders. We included both parallel‐group and cross‐over study designs. Data collection and analysis We used standard methodological procedures of Cochrane, in that two review authors independently selected studies, extracted data using standardized forms, assessed risk of bias, and graded the overall quality of the evidence by using the GRADE approach. We included eight randomized controlled trials (four of which were cross‐over trials) with 510 participants (443 boys, 67 girls) in this review. Participants in these studies were children with both ADHD and a chronic tic disorder. All studies took place in the USA and ranged from three to 22 weeks in duration. Five of the eight studies were funded by charitable organizations or government agencies, or both. One study was funded by the drug manufacturer. The other two studies did not specify the source of funding. Risk of bias of included studies was low for blinding; low or unclear for random sequence generation, allocation concealment, and attrition bias; and low or high for selective outcome reporting. We were unable to combine any of the studies in a meta‐analysis due to important clinical heterogeneity and unit‐of‐analysis issues. Several of the trials assessed multiple agents. Medications assessed included methylphenidate, clonidine, desipramine, dextroamphetamine, guanfacine, atomoxetine, and deprenyl. There was low‐quality evidence for methylphenidate, atomoxetine, and clonidine, and very low‐quality evidence for desipramine, dextroamphetamine, guanfacine and deprenyl in the treatment of ADHD in children with tics. All studies, with the exception of a study using deprenyl, reported improvement in symptoms of ADHD. Tic symptoms also improved in children treated with guanfacine, desipramine, methylphenidate, clonidine, and a combination of methylphenidate and clonidine. In one study, tics limited further dosage increases of methylphenidate. High‐dose dextroamphetamine appeared to worsen tics in one study, although the length of this study was limited to three weeks. There was appetite suppression or weight loss in association with methylphenidate, dextroamphetamine, atomoxetine, and desipramine. There was insomnia associated with methylphenidate and dextroamphetamine, and sedation associated with clonidine. Following an updated search of potentially relevant studies, we found no new studies that matched our inclusion criteria and thus our conclusions have not changed. Methylphenidate, clonidine, guanfacine, desipramine, and atomoxetine appear to reduce ADHD symptoms in children with tics though the quality of the available evidence was low to very low. Although stimulants have not been shown to worsen tics in most people with tic disorders, they may, nonetheless, exacerbate tics in individual cases. In these instances, treatment with alpha agonists or atomoxetine may be an alternative. Although there is evidence that desipramine may improve tics and ADHD in children, safety concerns will likely continue to limit its use in this population.
t141	t141_9	no	The other two studies did not specify the source of funding for the study.	This is an update of the original Cochrane Review published in Issue 4, 2011. Attention deficit hyperactivity disorder (ADHD) is the most prevalent of the comorbid psychiatric disorders that complicate tic disorders. Medications commonly used to treat ADHD symptoms include stimulants such as methylphenidate and amphetamine; non‐stimulants, such as atomoxetine; tricyclic antidepressants; and alpha agonists. Alpha agonists are also used as a treatment for tics. Due to the impact of ADHD symptoms on the child with tic disorder, treatment of ADHD is often of greater priority than the medical management of tics. However, for many decades, clinicians have been reluctant to use stimulants to treat children with ADHD and tics for fear of worsening their tics. Objectives To assess the effects of pharmacological treatments for ADHD in children with comorbid tic disorders on symptoms of ADHD and tics. Search methods In September 2017, we searched CENTRAL, MEDLINE, Embase, and 12 other databases. We also searched two trial registers and contacted experts in the field for any ongoing or unpublished studies. Selection criteria We included randomized, double‐blind, controlled trials of any pharmacological treatment for ADHD used specifically in children with comorbid tic disorders. We included both parallel‐group and cross‐over study designs. Data collection and analysis We used standard methodological procedures of Cochrane, in that two review authors independently selected studies, extracted data using standardized forms, assessed risk of bias, and graded the overall quality of the evidence by using the GRADE approach. We included eight randomized controlled trials (four of which were cross‐over trials) with 510 participants (443 boys, 67 girls) in this review. Participants in these studies were children with both ADHD and a chronic tic disorder. All studies took place in the USA and ranged from three to 22 weeks in duration. Five of the eight studies were funded by charitable organizations or government agencies, or both. One study was funded by the drug manufacturer. The other two studies did not specify the source of funding. Risk of bias of included studies was low for blinding; low or unclear for random sequence generation, allocation concealment, and attrition bias; and low or high for selective outcome reporting. We were unable to combine any of the studies in a meta‐analysis due to important clinical heterogeneity and unit‐of‐analysis issues. Several of the trials assessed multiple agents. Medications assessed included methylphenidate, clonidine, desipramine, dextroamphetamine, guanfacine, atomoxetine, and deprenyl. There was low‐quality evidence for methylphenidate, atomoxetine, and clonidine, and very low‐quality evidence for desipramine, dextroamphetamine, guanfacine and deprenyl in the treatment of ADHD in children with tics. All studies, with the exception of a study using deprenyl, reported improvement in symptoms of ADHD. Tic symptoms also improved in children treated with guanfacine, desipramine, methylphenidate, clonidine, and a combination of methylphenidate and clonidine. In one study, tics limited further dosage increases of methylphenidate. High‐dose dextroamphetamine appeared to worsen tics in one study, although the length of this study was limited to three weeks. There was appetite suppression or weight loss in association with methylphenidate, dextroamphetamine, atomoxetine, and desipramine. There was insomnia associated with methylphenidate and dextroamphetamine, and sedation associated with clonidine. Following an updated search of potentially relevant studies, we found no new studies that matched our inclusion criteria and thus our conclusions have not changed. Methylphenidate, clonidine, guanfacine, desipramine, and atomoxetine appear to reduce ADHD symptoms in children with tics though the quality of the available evidence was low to very low. Although stimulants have not been shown to worsen tics in most people with tic disorders, they may, nonetheless, exacerbate tics in individual cases. In these instances, treatment with alpha agonists or atomoxetine may be an alternative. Although there is evidence that desipramine may improve tics and ADHD in children, safety concerns will likely continue to limit its use in this population.
t141	t141_10	no	The trials in this review suggested that several stimulant and non‐stimulant medications may improve ADHD symptoms in children with ADHD and tics.	This is an update of the original Cochrane Review published in Issue 4, 2011. Attention deficit hyperactivity disorder (ADHD) is the most prevalent of the comorbid psychiatric disorders that complicate tic disorders. Medications commonly used to treat ADHD symptoms include stimulants such as methylphenidate and amphetamine; non‐stimulants, such as atomoxetine; tricyclic antidepressants; and alpha agonists. Alpha agonists are also used as a treatment for tics. Due to the impact of ADHD symptoms on the child with tic disorder, treatment of ADHD is often of greater priority than the medical management of tics. However, for many decades, clinicians have been reluctant to use stimulants to treat children with ADHD and tics for fear of worsening their tics. Objectives To assess the effects of pharmacological treatments for ADHD in children with comorbid tic disorders on symptoms of ADHD and tics. Search methods In September 2017, we searched CENTRAL, MEDLINE, Embase, and 12 other databases. We also searched two trial registers and contacted experts in the field for any ongoing or unpublished studies. Selection criteria We included randomized, double‐blind, controlled trials of any pharmacological treatment for ADHD used specifically in children with comorbid tic disorders. We included both parallel‐group and cross‐over study designs. Data collection and analysis We used standard methodological procedures of Cochrane, in that two review authors independently selected studies, extracted data using standardized forms, assessed risk of bias, and graded the overall quality of the evidence by using the GRADE approach. We included eight randomized controlled trials (four of which were cross‐over trials) with 510 participants (443 boys, 67 girls) in this review. Participants in these studies were children with both ADHD and a chronic tic disorder. All studies took place in the USA and ranged from three to 22 weeks in duration. Five of the eight studies were funded by charitable organizations or government agencies, or both. One study was funded by the drug manufacturer. The other two studies did not specify the source of funding. Risk of bias of included studies was low for blinding; low or unclear for random sequence generation, allocation concealment, and attrition bias; and low or high for selective outcome reporting. We were unable to combine any of the studies in a meta‐analysis due to important clinical heterogeneity and unit‐of‐analysis issues. Several of the trials assessed multiple agents. Medications assessed included methylphenidate, clonidine, desipramine, dextroamphetamine, guanfacine, atomoxetine, and deprenyl. There was low‐quality evidence for methylphenidate, atomoxetine, and clonidine, and very low‐quality evidence for desipramine, dextroamphetamine, guanfacine and deprenyl in the treatment of ADHD in children with tics. All studies, with the exception of a study using deprenyl, reported improvement in symptoms of ADHD. Tic symptoms also improved in children treated with guanfacine, desipramine, methylphenidate, clonidine, and a combination of methylphenidate and clonidine. In one study, tics limited further dosage increases of methylphenidate. High‐dose dextroamphetamine appeared to worsen tics in one study, although the length of this study was limited to three weeks. There was appetite suppression or weight loss in association with methylphenidate, dextroamphetamine, atomoxetine, and desipramine. There was insomnia associated with methylphenidate and dextroamphetamine, and sedation associated with clonidine. Following an updated search of potentially relevant studies, we found no new studies that matched our inclusion criteria and thus our conclusions have not changed. Methylphenidate, clonidine, guanfacine, desipramine, and atomoxetine appear to reduce ADHD symptoms in children with tics though the quality of the available evidence was low to very low. Although stimulants have not been shown to worsen tics in most people with tic disorders, they may, nonetheless, exacerbate tics in individual cases. In these instances, treatment with alpha agonists or atomoxetine may be an alternative. Although there is evidence that desipramine may improve tics and ADHD in children, safety concerns will likely continue to limit its use in this population.
t141	t141_11	no	At high doses, dextroamphetamine may initially worsen tics in some children, and dose increases of both dextroamphetamine and methylphenidate may be limited due to tic exacerbation.	This is an update of the original Cochrane Review published in Issue 4, 2011. Attention deficit hyperactivity disorder (ADHD) is the most prevalent of the comorbid psychiatric disorders that complicate tic disorders. Medications commonly used to treat ADHD symptoms include stimulants such as methylphenidate and amphetamine; non‐stimulants, such as atomoxetine; tricyclic antidepressants; and alpha agonists. Alpha agonists are also used as a treatment for tics. Due to the impact of ADHD symptoms on the child with tic disorder, treatment of ADHD is often of greater priority than the medical management of tics. However, for many decades, clinicians have been reluctant to use stimulants to treat children with ADHD and tics for fear of worsening their tics. Objectives To assess the effects of pharmacological treatments for ADHD in children with comorbid tic disorders on symptoms of ADHD and tics. Search methods In September 2017, we searched CENTRAL, MEDLINE, Embase, and 12 other databases. We also searched two trial registers and contacted experts in the field for any ongoing or unpublished studies. Selection criteria We included randomized, double‐blind, controlled trials of any pharmacological treatment for ADHD used specifically in children with comorbid tic disorders. We included both parallel‐group and cross‐over study designs. Data collection and analysis We used standard methodological procedures of Cochrane, in that two review authors independently selected studies, extracted data using standardized forms, assessed risk of bias, and graded the overall quality of the evidence by using the GRADE approach. We included eight randomized controlled trials (four of which were cross‐over trials) with 510 participants (443 boys, 67 girls) in this review. Participants in these studies were children with both ADHD and a chronic tic disorder. All studies took place in the USA and ranged from three to 22 weeks in duration. Five of the eight studies were funded by charitable organizations or government agencies, or both. One study was funded by the drug manufacturer. The other two studies did not specify the source of funding. Risk of bias of included studies was low for blinding; low or unclear for random sequence generation, allocation concealment, and attrition bias; and low or high for selective outcome reporting. We were unable to combine any of the studies in a meta‐analysis due to important clinical heterogeneity and unit‐of‐analysis issues. Several of the trials assessed multiple agents. Medications assessed included methylphenidate, clonidine, desipramine, dextroamphetamine, guanfacine, atomoxetine, and deprenyl. There was low‐quality evidence for methylphenidate, atomoxetine, and clonidine, and very low‐quality evidence for desipramine, dextroamphetamine, guanfacine and deprenyl in the treatment of ADHD in children with tics. All studies, with the exception of a study using deprenyl, reported improvement in symptoms of ADHD. Tic symptoms also improved in children treated with guanfacine, desipramine, methylphenidate, clonidine, and a combination of methylphenidate and clonidine. In one study, tics limited further dosage increases of methylphenidate. High‐dose dextroamphetamine appeared to worsen tics in one study, although the length of this study was limited to three weeks. There was appetite suppression or weight loss in association with methylphenidate, dextroamphetamine, atomoxetine, and desipramine. There was insomnia associated with methylphenidate and dextroamphetamine, and sedation associated with clonidine. Following an updated search of potentially relevant studies, we found no new studies that matched our inclusion criteria and thus our conclusions have not changed. Methylphenidate, clonidine, guanfacine, desipramine, and atomoxetine appear to reduce ADHD symptoms in children with tics though the quality of the available evidence was low to very low. Although stimulants have not been shown to worsen tics in most people with tic disorders, they may, nonetheless, exacerbate tics in individual cases. In these instances, treatment with alpha agonists or atomoxetine may be an alternative. Although there is evidence that desipramine may improve tics and ADHD in children, safety concerns will likely continue to limit its use in this population.
t141	t141_12	no	However, for most children, both tics and ADHD symptoms can improve with use of stimulant medications.	This is an update of the original Cochrane Review published in Issue 4, 2011. Attention deficit hyperactivity disorder (ADHD) is the most prevalent of the comorbid psychiatric disorders that complicate tic disorders. Medications commonly used to treat ADHD symptoms include stimulants such as methylphenidate and amphetamine; non‐stimulants, such as atomoxetine; tricyclic antidepressants; and alpha agonists. Alpha agonists are also used as a treatment for tics. Due to the impact of ADHD symptoms on the child with tic disorder, treatment of ADHD is often of greater priority than the medical management of tics. However, for many decades, clinicians have been reluctant to use stimulants to treat children with ADHD and tics for fear of worsening their tics. Objectives To assess the effects of pharmacological treatments for ADHD in children with comorbid tic disorders on symptoms of ADHD and tics. Search methods In September 2017, we searched CENTRAL, MEDLINE, Embase, and 12 other databases. We also searched two trial registers and contacted experts in the field for any ongoing or unpublished studies. Selection criteria We included randomized, double‐blind, controlled trials of any pharmacological treatment for ADHD used specifically in children with comorbid tic disorders. We included both parallel‐group and cross‐over study designs. Data collection and analysis We used standard methodological procedures of Cochrane, in that two review authors independently selected studies, extracted data using standardized forms, assessed risk of bias, and graded the overall quality of the evidence by using the GRADE approach. We included eight randomized controlled trials (four of which were cross‐over trials) with 510 participants (443 boys, 67 girls) in this review. Participants in these studies were children with both ADHD and a chronic tic disorder. All studies took place in the USA and ranged from three to 22 weeks in duration. Five of the eight studies were funded by charitable organizations or government agencies, or both. One study was funded by the drug manufacturer. The other two studies did not specify the source of funding. Risk of bias of included studies was low for blinding; low or unclear for random sequence generation, allocation concealment, and attrition bias; and low or high for selective outcome reporting. We were unable to combine any of the studies in a meta‐analysis due to important clinical heterogeneity and unit‐of‐analysis issues. Several of the trials assessed multiple agents. Medications assessed included methylphenidate, clonidine, desipramine, dextroamphetamine, guanfacine, atomoxetine, and deprenyl. There was low‐quality evidence for methylphenidate, atomoxetine, and clonidine, and very low‐quality evidence for desipramine, dextroamphetamine, guanfacine and deprenyl in the treatment of ADHD in children with tics. All studies, with the exception of a study using deprenyl, reported improvement in symptoms of ADHD. Tic symptoms also improved in children treated with guanfacine, desipramine, methylphenidate, clonidine, and a combination of methylphenidate and clonidine. In one study, tics limited further dosage increases of methylphenidate. High‐dose dextroamphetamine appeared to worsen tics in one study, although the length of this study was limited to three weeks. There was appetite suppression or weight loss in association with methylphenidate, dextroamphetamine, atomoxetine, and desipramine. There was insomnia associated with methylphenidate and dextroamphetamine, and sedation associated with clonidine. Following an updated search of potentially relevant studies, we found no new studies that matched our inclusion criteria and thus our conclusions have not changed. Methylphenidate, clonidine, guanfacine, desipramine, and atomoxetine appear to reduce ADHD symptoms in children with tics though the quality of the available evidence was low to very low. Although stimulants have not been shown to worsen tics in most people with tic disorders, they may, nonetheless, exacerbate tics in individual cases. In these instances, treatment with alpha agonists or atomoxetine may be an alternative. Although there is evidence that desipramine may improve tics and ADHD in children, safety concerns will likely continue to limit its use in this population.
t142	t142_1	yes	General anaesthetic reduces reflexes that stop regurgitated gastric juices reaching the lungs.	Fasting before general anaesthesia aims to reduce the volume and acidity of stomach contents during surgery, thus reducing the risk of regurgitation/aspiration. Recent guidelines have recommended a shift in fasting policy from the standard 'nil by mouth from midnight' approach to more relaxed policies which permit a period of restricted fluid intake up to a few hours before surgery. The evidence underpinning these guidelines however, was scattered across a range of journals, in a variety of languages, used a variety of outcome measures and methodologies to evaluate fasting regimens that differed in duration and the type and volume of intake permitted during a restricted fasting period. Practice has been slow to change. Objectives To systematically review the effect of different preoperative fasting regimens (duration, type and volume of permitted intake) on perioperative complications and patient wellbeing (including aspiration, regurgitation and related morbidity, thirst, hunger, pain, nausea, vomiting, anxiety) in different adult populations. Search methods Electronic databases, conference proceedings and reference lists from relevant articles were searched for studies of preoperative fasting in August 2003 and experts in the area were consulted. Selection criteria Randomised controlled trials which compared the effect on postoperative complications of different preoperative fasting regimens on adults were included. Data collection and analysis Details of the eligible studies were independently extracted by two reviewers and where relevant information was unavailable from the text attempts were made to contact the authors. Thirty eight randomised controlled comparisons (made within 22 trials) were identified. Most were based on 'healthy' adult participants who were not considered to be at increased risk of regurgitation or aspiration during anaesthesia. Few trials reported the incidence of aspiration/regurgitation or related morbidity but relied on indirect measures of patient safety i.e. intra‐operative gastric volume and pH. There was no evidence that the volume or pH of participants' gastric contents differed significantly depending on whether the groups were permitted a shortened preoperative fluid fast or continued a standard fast. Fluids evaluated included water, coffee, fruit juice, clear fluids and other drinks (e.g. isotonic drink, carbohydrate drink). Participants given a drink of water preoperatively were found to have a significantly lower volume of gastric contents than the groups that followed a standard fasting regimen. This difference was modest and clinically insignificant. There was no indication that the volume of fluid permitted during the preoperative period (i.e. low or high) resulted in a difference in outcomes from those participants that followed a standard fast. Few trials specifically investigated the preoperative fasting regimen for patient populations considered to be at increased risk during anaesthesia of regurgitation/aspiration and related morbidity. There was no evidence to suggest a shortened fluid fast results in an increased risk of aspiration, regurgitation or related morbidity compared with the standard 'nil by mouth from midnight' fasting policy. Permitting patients to drink water preoperatively resulted in significantly lower gastric volumes. Clinicians should be encouraged to appraise this evidence for themselves and when necessary adjust any remaining standard fasting policies (nil‐by‐mouth from midnight) for patients that are not considered 'at‐risk' during anaesthesia.
t142	t142_2	no	As this can be dangerous, people are often advised to have nothing to eat or drink from the midnight before surgery.	Fasting before general anaesthesia aims to reduce the volume and acidity of stomach contents during surgery, thus reducing the risk of regurgitation/aspiration. Recent guidelines have recommended a shift in fasting policy from the standard 'nil by mouth from midnight' approach to more relaxed policies which permit a period of restricted fluid intake up to a few hours before surgery. The evidence underpinning these guidelines however, was scattered across a range of journals, in a variety of languages, used a variety of outcome measures and methodologies to evaluate fasting regimens that differed in duration and the type and volume of intake permitted during a restricted fasting period. Practice has been slow to change. Objectives To systematically review the effect of different preoperative fasting regimens (duration, type and volume of permitted intake) on perioperative complications and patient wellbeing (including aspiration, regurgitation and related morbidity, thirst, hunger, pain, nausea, vomiting, anxiety) in different adult populations. Search methods Electronic databases, conference proceedings and reference lists from relevant articles were searched for studies of preoperative fasting in August 2003 and experts in the area were consulted. Selection criteria Randomised controlled trials which compared the effect on postoperative complications of different preoperative fasting regimens on adults were included. Data collection and analysis Details of the eligible studies were independently extracted by two reviewers and where relevant information was unavailable from the text attempts were made to contact the authors. Thirty eight randomised controlled comparisons (made within 22 trials) were identified. Most were based on 'healthy' adult participants who were not considered to be at increased risk of regurgitation or aspiration during anaesthesia. Few trials reported the incidence of aspiration/regurgitation or related morbidity but relied on indirect measures of patient safety i.e. intra‐operative gastric volume and pH. There was no evidence that the volume or pH of participants' gastric contents differed significantly depending on whether the groups were permitted a shortened preoperative fluid fast or continued a standard fast. Fluids evaluated included water, coffee, fruit juice, clear fluids and other drinks (e.g. isotonic drink, carbohydrate drink). Participants given a drink of water preoperatively were found to have a significantly lower volume of gastric contents than the groups that followed a standard fasting regimen. This difference was modest and clinically insignificant. There was no indication that the volume of fluid permitted during the preoperative period (i.e. low or high) resulted in a difference in outcomes from those participants that followed a standard fast. Few trials specifically investigated the preoperative fasting regimen for patient populations considered to be at increased risk during anaesthesia of regurgitation/aspiration and related morbidity. There was no evidence to suggest a shortened fluid fast results in an increased risk of aspiration, regurgitation or related morbidity compared with the standard 'nil by mouth from midnight' fasting policy. Permitting patients to drink water preoperatively resulted in significantly lower gastric volumes. Clinicians should be encouraged to appraise this evidence for themselves and when necessary adjust any remaining standard fasting policies (nil‐by‐mouth from midnight) for patients that are not considered 'at‐risk' during anaesthesia.
t142	t142_3	yes	However, the review of trials found that drinking clear fluids up to a few hours before surgery did not increase the risk of regurgitation during or after surgery.	Fasting before general anaesthesia aims to reduce the volume and acidity of stomach contents during surgery, thus reducing the risk of regurgitation/aspiration. Recent guidelines have recommended a shift in fasting policy from the standard 'nil by mouth from midnight' approach to more relaxed policies which permit a period of restricted fluid intake up to a few hours before surgery. The evidence underpinning these guidelines however, was scattered across a range of journals, in a variety of languages, used a variety of outcome measures and methodologies to evaluate fasting regimens that differed in duration and the type and volume of intake permitted during a restricted fasting period. Practice has been slow to change. Objectives To systematically review the effect of different preoperative fasting regimens (duration, type and volume of permitted intake) on perioperative complications and patient wellbeing (including aspiration, regurgitation and related morbidity, thirst, hunger, pain, nausea, vomiting, anxiety) in different adult populations. Search methods Electronic databases, conference proceedings and reference lists from relevant articles were searched for studies of preoperative fasting in August 2003 and experts in the area were consulted. Selection criteria Randomised controlled trials which compared the effect on postoperative complications of different preoperative fasting regimens on adults were included. Data collection and analysis Details of the eligible studies were independently extracted by two reviewers and where relevant information was unavailable from the text attempts were made to contact the authors. Thirty eight randomised controlled comparisons (made within 22 trials) were identified. Most were based on 'healthy' adult participants who were not considered to be at increased risk of regurgitation or aspiration during anaesthesia. Few trials reported the incidence of aspiration/regurgitation or related morbidity but relied on indirect measures of patient safety i.e. intra‐operative gastric volume and pH. There was no evidence that the volume or pH of participants' gastric contents differed significantly depending on whether the groups were permitted a shortened preoperative fluid fast or continued a standard fast. Fluids evaluated included water, coffee, fruit juice, clear fluids and other drinks (e.g. isotonic drink, carbohydrate drink). Participants given a drink of water preoperatively were found to have a significantly lower volume of gastric contents than the groups that followed a standard fasting regimen. This difference was modest and clinically insignificant. There was no indication that the volume of fluid permitted during the preoperative period (i.e. low or high) resulted in a difference in outcomes from those participants that followed a standard fast. Few trials specifically investigated the preoperative fasting regimen for patient populations considered to be at increased risk during anaesthesia of regurgitation/aspiration and related morbidity. There was no evidence to suggest a shortened fluid fast results in an increased risk of aspiration, regurgitation or related morbidity compared with the standard 'nil by mouth from midnight' fasting policy. Permitting patients to drink water preoperatively resulted in significantly lower gastric volumes. Clinicians should be encouraged to appraise this evidence for themselves and when necessary adjust any remaining standard fasting policies (nil‐by‐mouth from midnight) for patients that are not considered 'at‐risk' during anaesthesia.
t142	t142_4	yes	Some people are considered more likely to regurgitate under anaesthetic, including those who are pregnant, elderly, obese or have stomach disorders.	Fasting before general anaesthesia aims to reduce the volume and acidity of stomach contents during surgery, thus reducing the risk of regurgitation/aspiration. Recent guidelines have recommended a shift in fasting policy from the standard 'nil by mouth from midnight' approach to more relaxed policies which permit a period of restricted fluid intake up to a few hours before surgery. The evidence underpinning these guidelines however, was scattered across a range of journals, in a variety of languages, used a variety of outcome measures and methodologies to evaluate fasting regimens that differed in duration and the type and volume of intake permitted during a restricted fasting period. Practice has been slow to change. Objectives To systematically review the effect of different preoperative fasting regimens (duration, type and volume of permitted intake) on perioperative complications and patient wellbeing (including aspiration, regurgitation and related morbidity, thirst, hunger, pain, nausea, vomiting, anxiety) in different adult populations. Search methods Electronic databases, conference proceedings and reference lists from relevant articles were searched for studies of preoperative fasting in August 2003 and experts in the area were consulted. Selection criteria Randomised controlled trials which compared the effect on postoperative complications of different preoperative fasting regimens on adults were included. Data collection and analysis Details of the eligible studies were independently extracted by two reviewers and where relevant information was unavailable from the text attempts were made to contact the authors. Thirty eight randomised controlled comparisons (made within 22 trials) were identified. Most were based on 'healthy' adult participants who were not considered to be at increased risk of regurgitation or aspiration during anaesthesia. Few trials reported the incidence of aspiration/regurgitation or related morbidity but relied on indirect measures of patient safety i.e. intra‐operative gastric volume and pH. There was no evidence that the volume or pH of participants' gastric contents differed significantly depending on whether the groups were permitted a shortened preoperative fluid fast or continued a standard fast. Fluids evaluated included water, coffee, fruit juice, clear fluids and other drinks (e.g. isotonic drink, carbohydrate drink). Participants given a drink of water preoperatively were found to have a significantly lower volume of gastric contents than the groups that followed a standard fasting regimen. This difference was modest and clinically insignificant. There was no indication that the volume of fluid permitted during the preoperative period (i.e. low or high) resulted in a difference in outcomes from those participants that followed a standard fast. Few trials specifically investigated the preoperative fasting regimen for patient populations considered to be at increased risk during anaesthesia of regurgitation/aspiration and related morbidity. There was no evidence to suggest a shortened fluid fast results in an increased risk of aspiration, regurgitation or related morbidity compared with the standard 'nil by mouth from midnight' fasting policy. Permitting patients to drink water preoperatively resulted in significantly lower gastric volumes. Clinicians should be encouraged to appraise this evidence for themselves and when necessary adjust any remaining standard fasting policies (nil‐by‐mouth from midnight) for patients that are not considered 'at‐risk' during anaesthesia.
t142	t142_5	yes	More research is needed to determine whether these people can also safely drink up to a few hours before surgery.	Fasting before general anaesthesia aims to reduce the volume and acidity of stomach contents during surgery, thus reducing the risk of regurgitation/aspiration. Recent guidelines have recommended a shift in fasting policy from the standard 'nil by mouth from midnight' approach to more relaxed policies which permit a period of restricted fluid intake up to a few hours before surgery. The evidence underpinning these guidelines however, was scattered across a range of journals, in a variety of languages, used a variety of outcome measures and methodologies to evaluate fasting regimens that differed in duration and the type and volume of intake permitted during a restricted fasting period. Practice has been slow to change. Objectives To systematically review the effect of different preoperative fasting regimens (duration, type and volume of permitted intake) on perioperative complications and patient wellbeing (including aspiration, regurgitation and related morbidity, thirst, hunger, pain, nausea, vomiting, anxiety) in different adult populations. Search methods Electronic databases, conference proceedings and reference lists from relevant articles were searched for studies of preoperative fasting in August 2003 and experts in the area were consulted. Selection criteria Randomised controlled trials which compared the effect on postoperative complications of different preoperative fasting regimens on adults were included. Data collection and analysis Details of the eligible studies were independently extracted by two reviewers and where relevant information was unavailable from the text attempts were made to contact the authors. Thirty eight randomised controlled comparisons (made within 22 trials) were identified. Most were based on 'healthy' adult participants who were not considered to be at increased risk of regurgitation or aspiration during anaesthesia. Few trials reported the incidence of aspiration/regurgitation or related morbidity but relied on indirect measures of patient safety i.e. intra‐operative gastric volume and pH. There was no evidence that the volume or pH of participants' gastric contents differed significantly depending on whether the groups were permitted a shortened preoperative fluid fast or continued a standard fast. Fluids evaluated included water, coffee, fruit juice, clear fluids and other drinks (e.g. isotonic drink, carbohydrate drink). Participants given a drink of water preoperatively were found to have a significantly lower volume of gastric contents than the groups that followed a standard fasting regimen. This difference was modest and clinically insignificant. There was no indication that the volume of fluid permitted during the preoperative period (i.e. low or high) resulted in a difference in outcomes from those participants that followed a standard fast. Few trials specifically investigated the preoperative fasting regimen for patient populations considered to be at increased risk during anaesthesia of regurgitation/aspiration and related morbidity. There was no evidence to suggest a shortened fluid fast results in an increased risk of aspiration, regurgitation or related morbidity compared with the standard 'nil by mouth from midnight' fasting policy. Permitting patients to drink water preoperatively resulted in significantly lower gastric volumes. Clinicians should be encouraged to appraise this evidence for themselves and when necessary adjust any remaining standard fasting policies (nil‐by‐mouth from midnight) for patients that are not considered 'at‐risk' during anaesthesia.
t143	t143_1	no	There is much debate on the diagnostic performance of endoscopic ultrasound (EUS) in the preoperative staging of gastric cancer.	Endoscopic ultrasound (EUS) is proposed as an accurate diagnostic device for the locoregional staging of gastric cancer, which is crucial to developing a correct therapeutic strategy and ultimately to providing patients with the best chance of cure. However, despite a number of studies addressing this issue, there is no consensus on the role of EUS in routine clinical practice. Objectives To provide both a comprehensive overview and a quantitative analysis of the published data regarding the ability of EUS to preoperatively define the locoregional disease spread (i.e., primary tumor depth (T‐stage) and regional lymph node status (N‐stage)) in people with primary gastric carcinoma. Search methods We performed a systematic search to identify articles that examined the diagnostic accuracy of EUS (the index test) in the evaluation of primary gastric cancer depth of invasion (T‐stage, according to the AJCC/UICC TNM staging system categories T1, T2, T3 and T4) and regional lymph node status (N‐stage, disease‐free (N0) versus metastatic (N+)) using histopathology as the reference standard. To this end, we searched the following databases: the Cochrane Library (the Cochrane Central Register of Controlled Trials (CENTRAL)), MEDLINE, EMBASE , NIHR Prospero Register, MEDION, Aggressive Research Intelligence Facility (ARIF), ClinicalTrials.gov, Current Controlled Trials MetaRegister, and World Health Organization International Clinical Trials Registry Platform (WHO ICTRP), from 1988 to January 2015. Selection criteria We included studies that met the following main inclusion criteria: 1) a minimum sample size of 10 patients with histologically‐proven primary carcinoma of the stomach (target condition); 2) comparison of EUS (index test) with pathology evaluation (reference standard) in terms of primary tumor (T‐stage) and regional lymph nodes (N‐stage). We excluded reports with possible overlap with the selected studies. Data collection and analysis For each study, two review authors extracted a standard set of data, using a dedicated data extraction form. We assessed data quality using a standard procedure according to the Quality Assessment of Diagnostic Accuracy Studies (QUADAS‐2) criteria. We performed diagnostic accuracy meta‐analysis using the hierarchical bivariate method. We identified 66 articles (published between 1988 and 2012) that were eligible according to the inclusion criteria. We collected the data on 7747 patients with gastric cancer who were staged with EUS. Overall the quality of the included studies was good: in particular, only five studies presented a high risk of index test interpretation bias and two studies presented a high risk of selection bias. For primary tumor (T) stage, results were stratified according to the depth of invasion of the gastric wall. The meta‐analysis of 50 studies (n = 4397) showed that the summary sensitivity and specificity of EUS in discriminating T1 to T2 (superficial) versus T3 to T4 (advanced) gastric carcinomas were 0.86 (95% confidence interval (CI) 0.81 to 0.90) and 0.90 (95% CI 0.87 to 0.93) respectively. For the diagnostic capacity of EUS to distinguish T1 (early gastric cancer, EGC) versus T2 (muscle‐infiltrating) tumors, the meta‐analysis of 46 studies (n = 2742) showed that the summary sensitivity and specificity were 0.85 (95% CI 0.78 to 0.91) and 0.90 (95% CI 0.85 to 0.93) respectively. When we addressed the capacity of EUS to distinguish between T1a (mucosal) versus T1b (submucosal) cancers the meta‐analysis of 20 studies (n = 3321) showed that the summary sensitivity and specificity were 0.87 (95% CI 0.81 to 0.92) and 0.75 (95% CI 0.62 to 0.84) respectively. Finally, for the metastatic involvement of lymph nodes (N‐stage), the meta‐analysis of 44 studies (n = 3573) showed that the summary sensitivity and specificity were 0.83 (95% CI 0.79 to 0.87) and 0.67 (95% CI 0.61 to 0.72), respectively. Overall, as demonstrated also by the Bayesian nomograms, which enable readers to calculate post‐test probabilities for any target condition prevalence, the EUS accuracy can be considered clinically useful to guide physicians in the locoregional staging of people with gastric cancer. However, it should be noted that between‐study heterogeneity was not negligible: unfortunately, we could not identify any consistent source of the observed heterogeneity. Therefore, all accuracy measures reported in the present work and summarizing the available evidence should be interpreted cautiously. Moreover, we must emphasize that the analysis of positive and negative likelihood values revealed that EUS diagnostic performance cannot be considered optimal either for disease confirmation or for exclusion, especially for the ability of EUS to distinguish T1a (mucosal) versus T1b (submucosal) cancers and positive versus negative lymph node status. By analyzing the data from the largest series ever considered, we found that the diagnostic accuracy of EUS might be considered clinically useful to guide physicians in the locoregional staging of people with gastric carcinoma. However, the heterogeneity of the results warrants special caution, as well as further investigation for the identification of factors influencing the outcome of this diagnostic tool. Moreover, physicians should be warned that EUS performance is lower in diagnosing superficial tumors (T1a versus T1b) and lymph node status (positive versus negative). Overall, we observed large heterogeneity and its source needs to be understood before any definitive conclusion can be drawn about the use of EUS can be proposed in routine clinical settings.
t143	t143_2	no	The aim of this review was to collect the available evidence and then to calculate how well EUS stages stomach cancer.	Endoscopic ultrasound (EUS) is proposed as an accurate diagnostic device for the locoregional staging of gastric cancer, which is crucial to developing a correct therapeutic strategy and ultimately to providing patients with the best chance of cure. However, despite a number of studies addressing this issue, there is no consensus on the role of EUS in routine clinical practice. Objectives To provide both a comprehensive overview and a quantitative analysis of the published data regarding the ability of EUS to preoperatively define the locoregional disease spread (i.e., primary tumor depth (T‐stage) and regional lymph node status (N‐stage)) in people with primary gastric carcinoma. Search methods We performed a systematic search to identify articles that examined the diagnostic accuracy of EUS (the index test) in the evaluation of primary gastric cancer depth of invasion (T‐stage, according to the AJCC/UICC TNM staging system categories T1, T2, T3 and T4) and regional lymph node status (N‐stage, disease‐free (N0) versus metastatic (N+)) using histopathology as the reference standard. To this end, we searched the following databases: the Cochrane Library (the Cochrane Central Register of Controlled Trials (CENTRAL)), MEDLINE, EMBASE , NIHR Prospero Register, MEDION, Aggressive Research Intelligence Facility (ARIF), ClinicalTrials.gov, Current Controlled Trials MetaRegister, and World Health Organization International Clinical Trials Registry Platform (WHO ICTRP), from 1988 to January 2015. Selection criteria We included studies that met the following main inclusion criteria: 1) a minimum sample size of 10 patients with histologically‐proven primary carcinoma of the stomach (target condition); 2) comparison of EUS (index test) with pathology evaluation (reference standard) in terms of primary tumor (T‐stage) and regional lymph nodes (N‐stage). We excluded reports with possible overlap with the selected studies. Data collection and analysis For each study, two review authors extracted a standard set of data, using a dedicated data extraction form. We assessed data quality using a standard procedure according to the Quality Assessment of Diagnostic Accuracy Studies (QUADAS‐2) criteria. We performed diagnostic accuracy meta‐analysis using the hierarchical bivariate method. We identified 66 articles (published between 1988 and 2012) that were eligible according to the inclusion criteria. We collected the data on 7747 patients with gastric cancer who were staged with EUS. Overall the quality of the included studies was good: in particular, only five studies presented a high risk of index test interpretation bias and two studies presented a high risk of selection bias. For primary tumor (T) stage, results were stratified according to the depth of invasion of the gastric wall. The meta‐analysis of 50 studies (n = 4397) showed that the summary sensitivity and specificity of EUS in discriminating T1 to T2 (superficial) versus T3 to T4 (advanced) gastric carcinomas were 0.86 (95% confidence interval (CI) 0.81 to 0.90) and 0.90 (95% CI 0.87 to 0.93) respectively. For the diagnostic capacity of EUS to distinguish T1 (early gastric cancer, EGC) versus T2 (muscle‐infiltrating) tumors, the meta‐analysis of 46 studies (n = 2742) showed that the summary sensitivity and specificity were 0.85 (95% CI 0.78 to 0.91) and 0.90 (95% CI 0.85 to 0.93) respectively. When we addressed the capacity of EUS to distinguish between T1a (mucosal) versus T1b (submucosal) cancers the meta‐analysis of 20 studies (n = 3321) showed that the summary sensitivity and specificity were 0.87 (95% CI 0.81 to 0.92) and 0.75 (95% CI 0.62 to 0.84) respectively. Finally, for the metastatic involvement of lymph nodes (N‐stage), the meta‐analysis of 44 studies (n = 3573) showed that the summary sensitivity and specificity were 0.83 (95% CI 0.79 to 0.87) and 0.67 (95% CI 0.61 to 0.72), respectively. Overall, as demonstrated also by the Bayesian nomograms, which enable readers to calculate post‐test probabilities for any target condition prevalence, the EUS accuracy can be considered clinically useful to guide physicians in the locoregional staging of people with gastric cancer. However, it should be noted that between‐study heterogeneity was not negligible: unfortunately, we could not identify any consistent source of the observed heterogeneity. Therefore, all accuracy measures reported in the present work and summarizing the available evidence should be interpreted cautiously. Moreover, we must emphasize that the analysis of positive and negative likelihood values revealed that EUS diagnostic performance cannot be considered optimal either for disease confirmation or for exclusion, especially for the ability of EUS to distinguish T1a (mucosal) versus T1b (submucosal) cancers and positive versus negative lymph node status. By analyzing the data from the largest series ever considered, we found that the diagnostic accuracy of EUS might be considered clinically useful to guide physicians in the locoregional staging of people with gastric carcinoma. However, the heterogeneity of the results warrants special caution, as well as further investigation for the identification of factors influencing the outcome of this diagnostic tool. Moreover, physicians should be warned that EUS performance is lower in diagnosing superficial tumors (T1a versus T1b) and lymph node status (positive versus negative). Overall, we observed large heterogeneity and its source needs to be understood before any definitive conclusion can be drawn about the use of EUS can be proposed in routine clinical settings.
t143	t143_3	no	EUS is a diagnostic test that can be used to determine how far (stage) cancer of the stomach reaches prior to surgery.	Endoscopic ultrasound (EUS) is proposed as an accurate diagnostic device for the locoregional staging of gastric cancer, which is crucial to developing a correct therapeutic strategy and ultimately to providing patients with the best chance of cure. However, despite a number of studies addressing this issue, there is no consensus on the role of EUS in routine clinical practice. Objectives To provide both a comprehensive overview and a quantitative analysis of the published data regarding the ability of EUS to preoperatively define the locoregional disease spread (i.e., primary tumor depth (T‐stage) and regional lymph node status (N‐stage)) in people with primary gastric carcinoma. Search methods We performed a systematic search to identify articles that examined the diagnostic accuracy of EUS (the index test) in the evaluation of primary gastric cancer depth of invasion (T‐stage, according to the AJCC/UICC TNM staging system categories T1, T2, T3 and T4) and regional lymph node status (N‐stage, disease‐free (N0) versus metastatic (N+)) using histopathology as the reference standard. To this end, we searched the following databases: the Cochrane Library (the Cochrane Central Register of Controlled Trials (CENTRAL)), MEDLINE, EMBASE , NIHR Prospero Register, MEDION, Aggressive Research Intelligence Facility (ARIF), ClinicalTrials.gov, Current Controlled Trials MetaRegister, and World Health Organization International Clinical Trials Registry Platform (WHO ICTRP), from 1988 to January 2015. Selection criteria We included studies that met the following main inclusion criteria: 1) a minimum sample size of 10 patients with histologically‐proven primary carcinoma of the stomach (target condition); 2) comparison of EUS (index test) with pathology evaluation (reference standard) in terms of primary tumor (T‐stage) and regional lymph nodes (N‐stage). We excluded reports with possible overlap with the selected studies. Data collection and analysis For each study, two review authors extracted a standard set of data, using a dedicated data extraction form. We assessed data quality using a standard procedure according to the Quality Assessment of Diagnostic Accuracy Studies (QUADAS‐2) criteria. We performed diagnostic accuracy meta‐analysis using the hierarchical bivariate method. We identified 66 articles (published between 1988 and 2012) that were eligible according to the inclusion criteria. We collected the data on 7747 patients with gastric cancer who were staged with EUS. Overall the quality of the included studies was good: in particular, only five studies presented a high risk of index test interpretation bias and two studies presented a high risk of selection bias. For primary tumor (T) stage, results were stratified according to the depth of invasion of the gastric wall. The meta‐analysis of 50 studies (n = 4397) showed that the summary sensitivity and specificity of EUS in discriminating T1 to T2 (superficial) versus T3 to T4 (advanced) gastric carcinomas were 0.86 (95% confidence interval (CI) 0.81 to 0.90) and 0.90 (95% CI 0.87 to 0.93) respectively. For the diagnostic capacity of EUS to distinguish T1 (early gastric cancer, EGC) versus T2 (muscle‐infiltrating) tumors, the meta‐analysis of 46 studies (n = 2742) showed that the summary sensitivity and specificity were 0.85 (95% CI 0.78 to 0.91) and 0.90 (95% CI 0.85 to 0.93) respectively. When we addressed the capacity of EUS to distinguish between T1a (mucosal) versus T1b (submucosal) cancers the meta‐analysis of 20 studies (n = 3321) showed that the summary sensitivity and specificity were 0.87 (95% CI 0.81 to 0.92) and 0.75 (95% CI 0.62 to 0.84) respectively. Finally, for the metastatic involvement of lymph nodes (N‐stage), the meta‐analysis of 44 studies (n = 3573) showed that the summary sensitivity and specificity were 0.83 (95% CI 0.79 to 0.87) and 0.67 (95% CI 0.61 to 0.72), respectively. Overall, as demonstrated also by the Bayesian nomograms, which enable readers to calculate post‐test probabilities for any target condition prevalence, the EUS accuracy can be considered clinically useful to guide physicians in the locoregional staging of people with gastric cancer. However, it should be noted that between‐study heterogeneity was not negligible: unfortunately, we could not identify any consistent source of the observed heterogeneity. Therefore, all accuracy measures reported in the present work and summarizing the available evidence should be interpreted cautiously. Moreover, we must emphasize that the analysis of positive and negative likelihood values revealed that EUS diagnostic performance cannot be considered optimal either for disease confirmation or for exclusion, especially for the ability of EUS to distinguish T1a (mucosal) versus T1b (submucosal) cancers and positive versus negative lymph node status. By analyzing the data from the largest series ever considered, we found that the diagnostic accuracy of EUS might be considered clinically useful to guide physicians in the locoregional staging of people with gastric carcinoma. However, the heterogeneity of the results warrants special caution, as well as further investigation for the identification of factors influencing the outcome of this diagnostic tool. Moreover, physicians should be warned that EUS performance is lower in diagnosing superficial tumors (T1a versus T1b) and lymph node status (positive versus negative). Overall, we observed large heterogeneity and its source needs to be understood before any definitive conclusion can be drawn about the use of EUS can be proposed in routine clinical settings.
t143	t143_4	yes	It consists of an endoscope coupled with an ultrasound device capable of scanning the stomach wall, which shows the different layers of the stomach.	Endoscopic ultrasound (EUS) is proposed as an accurate diagnostic device for the locoregional staging of gastric cancer, which is crucial to developing a correct therapeutic strategy and ultimately to providing patients with the best chance of cure. However, despite a number of studies addressing this issue, there is no consensus on the role of EUS in routine clinical practice. Objectives To provide both a comprehensive overview and a quantitative analysis of the published data regarding the ability of EUS to preoperatively define the locoregional disease spread (i.e., primary tumor depth (T‐stage) and regional lymph node status (N‐stage)) in people with primary gastric carcinoma. Search methods We performed a systematic search to identify articles that examined the diagnostic accuracy of EUS (the index test) in the evaluation of primary gastric cancer depth of invasion (T‐stage, according to the AJCC/UICC TNM staging system categories T1, T2, T3 and T4) and regional lymph node status (N‐stage, disease‐free (N0) versus metastatic (N+)) using histopathology as the reference standard. To this end, we searched the following databases: the Cochrane Library (the Cochrane Central Register of Controlled Trials (CENTRAL)), MEDLINE, EMBASE , NIHR Prospero Register, MEDION, Aggressive Research Intelligence Facility (ARIF), ClinicalTrials.gov, Current Controlled Trials MetaRegister, and World Health Organization International Clinical Trials Registry Platform (WHO ICTRP), from 1988 to January 2015. Selection criteria We included studies that met the following main inclusion criteria: 1) a minimum sample size of 10 patients with histologically‐proven primary carcinoma of the stomach (target condition); 2) comparison of EUS (index test) with pathology evaluation (reference standard) in terms of primary tumor (T‐stage) and regional lymph nodes (N‐stage). We excluded reports with possible overlap with the selected studies. Data collection and analysis For each study, two review authors extracted a standard set of data, using a dedicated data extraction form. We assessed data quality using a standard procedure according to the Quality Assessment of Diagnostic Accuracy Studies (QUADAS‐2) criteria. We performed diagnostic accuracy meta‐analysis using the hierarchical bivariate method. We identified 66 articles (published between 1988 and 2012) that were eligible according to the inclusion criteria. We collected the data on 7747 patients with gastric cancer who were staged with EUS. Overall the quality of the included studies was good: in particular, only five studies presented a high risk of index test interpretation bias and two studies presented a high risk of selection bias. For primary tumor (T) stage, results were stratified according to the depth of invasion of the gastric wall. The meta‐analysis of 50 studies (n = 4397) showed that the summary sensitivity and specificity of EUS in discriminating T1 to T2 (superficial) versus T3 to T4 (advanced) gastric carcinomas were 0.86 (95% confidence interval (CI) 0.81 to 0.90) and 0.90 (95% CI 0.87 to 0.93) respectively. For the diagnostic capacity of EUS to distinguish T1 (early gastric cancer, EGC) versus T2 (muscle‐infiltrating) tumors, the meta‐analysis of 46 studies (n = 2742) showed that the summary sensitivity and specificity were 0.85 (95% CI 0.78 to 0.91) and 0.90 (95% CI 0.85 to 0.93) respectively. When we addressed the capacity of EUS to distinguish between T1a (mucosal) versus T1b (submucosal) cancers the meta‐analysis of 20 studies (n = 3321) showed that the summary sensitivity and specificity were 0.87 (95% CI 0.81 to 0.92) and 0.75 (95% CI 0.62 to 0.84) respectively. Finally, for the metastatic involvement of lymph nodes (N‐stage), the meta‐analysis of 44 studies (n = 3573) showed that the summary sensitivity and specificity were 0.83 (95% CI 0.79 to 0.87) and 0.67 (95% CI 0.61 to 0.72), respectively. Overall, as demonstrated also by the Bayesian nomograms, which enable readers to calculate post‐test probabilities for any target condition prevalence, the EUS accuracy can be considered clinically useful to guide physicians in the locoregional staging of people with gastric cancer. However, it should be noted that between‐study heterogeneity was not negligible: unfortunately, we could not identify any consistent source of the observed heterogeneity. Therefore, all accuracy measures reported in the present work and summarizing the available evidence should be interpreted cautiously. Moreover, we must emphasize that the analysis of positive and negative likelihood values revealed that EUS diagnostic performance cannot be considered optimal either for disease confirmation or for exclusion, especially for the ability of EUS to distinguish T1a (mucosal) versus T1b (submucosal) cancers and positive versus negative lymph node status. By analyzing the data from the largest series ever considered, we found that the diagnostic accuracy of EUS might be considered clinically useful to guide physicians in the locoregional staging of people with gastric carcinoma. However, the heterogeneity of the results warrants special caution, as well as further investigation for the identification of factors influencing the outcome of this diagnostic tool. Moreover, physicians should be warned that EUS performance is lower in diagnosing superficial tumors (T1a versus T1b) and lymph node status (positive versus negative). Overall, we observed large heterogeneity and its source needs to be understood before any definitive conclusion can be drawn about the use of EUS can be proposed in routine clinical settings.
t143	t143_5	yes	Changes from the normal ultrasonographic patterns due to the tumor growth can be used to determine the extent of cancer in the stomach wall (T‐stage) and the lymph nodes related to the stomach (N‐stage).	Endoscopic ultrasound (EUS) is proposed as an accurate diagnostic device for the locoregional staging of gastric cancer, which is crucial to developing a correct therapeutic strategy and ultimately to providing patients with the best chance of cure. However, despite a number of studies addressing this issue, there is no consensus on the role of EUS in routine clinical practice. Objectives To provide both a comprehensive overview and a quantitative analysis of the published data regarding the ability of EUS to preoperatively define the locoregional disease spread (i.e., primary tumor depth (T‐stage) and regional lymph node status (N‐stage)) in people with primary gastric carcinoma. Search methods We performed a systematic search to identify articles that examined the diagnostic accuracy of EUS (the index test) in the evaluation of primary gastric cancer depth of invasion (T‐stage, according to the AJCC/UICC TNM staging system categories T1, T2, T3 and T4) and regional lymph node status (N‐stage, disease‐free (N0) versus metastatic (N+)) using histopathology as the reference standard. To this end, we searched the following databases: the Cochrane Library (the Cochrane Central Register of Controlled Trials (CENTRAL)), MEDLINE, EMBASE , NIHR Prospero Register, MEDION, Aggressive Research Intelligence Facility (ARIF), ClinicalTrials.gov, Current Controlled Trials MetaRegister, and World Health Organization International Clinical Trials Registry Platform (WHO ICTRP), from 1988 to January 2015. Selection criteria We included studies that met the following main inclusion criteria: 1) a minimum sample size of 10 patients with histologically‐proven primary carcinoma of the stomach (target condition); 2) comparison of EUS (index test) with pathology evaluation (reference standard) in terms of primary tumor (T‐stage) and regional lymph nodes (N‐stage). We excluded reports with possible overlap with the selected studies. Data collection and analysis For each study, two review authors extracted a standard set of data, using a dedicated data extraction form. We assessed data quality using a standard procedure according to the Quality Assessment of Diagnostic Accuracy Studies (QUADAS‐2) criteria. We performed diagnostic accuracy meta‐analysis using the hierarchical bivariate method. We identified 66 articles (published between 1988 and 2012) that were eligible according to the inclusion criteria. We collected the data on 7747 patients with gastric cancer who were staged with EUS. Overall the quality of the included studies was good: in particular, only five studies presented a high risk of index test interpretation bias and two studies presented a high risk of selection bias. For primary tumor (T) stage, results were stratified according to the depth of invasion of the gastric wall. The meta‐analysis of 50 studies (n = 4397) showed that the summary sensitivity and specificity of EUS in discriminating T1 to T2 (superficial) versus T3 to T4 (advanced) gastric carcinomas were 0.86 (95% confidence interval (CI) 0.81 to 0.90) and 0.90 (95% CI 0.87 to 0.93) respectively. For the diagnostic capacity of EUS to distinguish T1 (early gastric cancer, EGC) versus T2 (muscle‐infiltrating) tumors, the meta‐analysis of 46 studies (n = 2742) showed that the summary sensitivity and specificity were 0.85 (95% CI 0.78 to 0.91) and 0.90 (95% CI 0.85 to 0.93) respectively. When we addressed the capacity of EUS to distinguish between T1a (mucosal) versus T1b (submucosal) cancers the meta‐analysis of 20 studies (n = 3321) showed that the summary sensitivity and specificity were 0.87 (95% CI 0.81 to 0.92) and 0.75 (95% CI 0.62 to 0.84) respectively. Finally, for the metastatic involvement of lymph nodes (N‐stage), the meta‐analysis of 44 studies (n = 3573) showed that the summary sensitivity and specificity were 0.83 (95% CI 0.79 to 0.87) and 0.67 (95% CI 0.61 to 0.72), respectively. Overall, as demonstrated also by the Bayesian nomograms, which enable readers to calculate post‐test probabilities for any target condition prevalence, the EUS accuracy can be considered clinically useful to guide physicians in the locoregional staging of people with gastric cancer. However, it should be noted that between‐study heterogeneity was not negligible: unfortunately, we could not identify any consistent source of the observed heterogeneity. Therefore, all accuracy measures reported in the present work and summarizing the available evidence should be interpreted cautiously. Moreover, we must emphasize that the analysis of positive and negative likelihood values revealed that EUS diagnostic performance cannot be considered optimal either for disease confirmation or for exclusion, especially for the ability of EUS to distinguish T1a (mucosal) versus T1b (submucosal) cancers and positive versus negative lymph node status. By analyzing the data from the largest series ever considered, we found that the diagnostic accuracy of EUS might be considered clinically useful to guide physicians in the locoregional staging of people with gastric carcinoma. However, the heterogeneity of the results warrants special caution, as well as further investigation for the identification of factors influencing the outcome of this diagnostic tool. Moreover, physicians should be warned that EUS performance is lower in diagnosing superficial tumors (T1a versus T1b) and lymph node status (positive versus negative). Overall, we observed large heterogeneity and its source needs to be understood before any definitive conclusion can be drawn about the use of EUS can be proposed in routine clinical settings.
t143	t143_6	no	Since the correct staging of the tumor enables physicians to personalize cancer treatment, it is important to understand the reliability of staging devices.	Endoscopic ultrasound (EUS) is proposed as an accurate diagnostic device for the locoregional staging of gastric cancer, which is crucial to developing a correct therapeutic strategy and ultimately to providing patients with the best chance of cure. However, despite a number of studies addressing this issue, there is no consensus on the role of EUS in routine clinical practice. Objectives To provide both a comprehensive overview and a quantitative analysis of the published data regarding the ability of EUS to preoperatively define the locoregional disease spread (i.e., primary tumor depth (T‐stage) and regional lymph node status (N‐stage)) in people with primary gastric carcinoma. Search methods We performed a systematic search to identify articles that examined the diagnostic accuracy of EUS (the index test) in the evaluation of primary gastric cancer depth of invasion (T‐stage, according to the AJCC/UICC TNM staging system categories T1, T2, T3 and T4) and regional lymph node status (N‐stage, disease‐free (N0) versus metastatic (N+)) using histopathology as the reference standard. To this end, we searched the following databases: the Cochrane Library (the Cochrane Central Register of Controlled Trials (CENTRAL)), MEDLINE, EMBASE , NIHR Prospero Register, MEDION, Aggressive Research Intelligence Facility (ARIF), ClinicalTrials.gov, Current Controlled Trials MetaRegister, and World Health Organization International Clinical Trials Registry Platform (WHO ICTRP), from 1988 to January 2015. Selection criteria We included studies that met the following main inclusion criteria: 1) a minimum sample size of 10 patients with histologically‐proven primary carcinoma of the stomach (target condition); 2) comparison of EUS (index test) with pathology evaluation (reference standard) in terms of primary tumor (T‐stage) and regional lymph nodes (N‐stage). We excluded reports with possible overlap with the selected studies. Data collection and analysis For each study, two review authors extracted a standard set of data, using a dedicated data extraction form. We assessed data quality using a standard procedure according to the Quality Assessment of Diagnostic Accuracy Studies (QUADAS‐2) criteria. We performed diagnostic accuracy meta‐analysis using the hierarchical bivariate method. We identified 66 articles (published between 1988 and 2012) that were eligible according to the inclusion criteria. We collected the data on 7747 patients with gastric cancer who were staged with EUS. Overall the quality of the included studies was good: in particular, only five studies presented a high risk of index test interpretation bias and two studies presented a high risk of selection bias. For primary tumor (T) stage, results were stratified according to the depth of invasion of the gastric wall. The meta‐analysis of 50 studies (n = 4397) showed that the summary sensitivity and specificity of EUS in discriminating T1 to T2 (superficial) versus T3 to T4 (advanced) gastric carcinomas were 0.86 (95% confidence interval (CI) 0.81 to 0.90) and 0.90 (95% CI 0.87 to 0.93) respectively. For the diagnostic capacity of EUS to distinguish T1 (early gastric cancer, EGC) versus T2 (muscle‐infiltrating) tumors, the meta‐analysis of 46 studies (n = 2742) showed that the summary sensitivity and specificity were 0.85 (95% CI 0.78 to 0.91) and 0.90 (95% CI 0.85 to 0.93) respectively. When we addressed the capacity of EUS to distinguish between T1a (mucosal) versus T1b (submucosal) cancers the meta‐analysis of 20 studies (n = 3321) showed that the summary sensitivity and specificity were 0.87 (95% CI 0.81 to 0.92) and 0.75 (95% CI 0.62 to 0.84) respectively. Finally, for the metastatic involvement of lymph nodes (N‐stage), the meta‐analysis of 44 studies (n = 3573) showed that the summary sensitivity and specificity were 0.83 (95% CI 0.79 to 0.87) and 0.67 (95% CI 0.61 to 0.72), respectively. Overall, as demonstrated also by the Bayesian nomograms, which enable readers to calculate post‐test probabilities for any target condition prevalence, the EUS accuracy can be considered clinically useful to guide physicians in the locoregional staging of people with gastric cancer. However, it should be noted that between‐study heterogeneity was not negligible: unfortunately, we could not identify any consistent source of the observed heterogeneity. Therefore, all accuracy measures reported in the present work and summarizing the available evidence should be interpreted cautiously. Moreover, we must emphasize that the analysis of positive and negative likelihood values revealed that EUS diagnostic performance cannot be considered optimal either for disease confirmation or for exclusion, especially for the ability of EUS to distinguish T1a (mucosal) versus T1b (submucosal) cancers and positive versus negative lymph node status. By analyzing the data from the largest series ever considered, we found that the diagnostic accuracy of EUS might be considered clinically useful to guide physicians in the locoregional staging of people with gastric carcinoma. However, the heterogeneity of the results warrants special caution, as well as further investigation for the identification of factors influencing the outcome of this diagnostic tool. Moreover, physicians should be warned that EUS performance is lower in diagnosing superficial tumors (T1a versus T1b) and lymph node status (positive versus negative). Overall, we observed large heterogeneity and its source needs to be understood before any definitive conclusion can be drawn about the use of EUS can be proposed in routine clinical settings.
t143	t143_7	yes	We conducted a meta‐analysis according to the most recent methods for diagnostic tests.	Endoscopic ultrasound (EUS) is proposed as an accurate diagnostic device for the locoregional staging of gastric cancer, which is crucial to developing a correct therapeutic strategy and ultimately to providing patients with the best chance of cure. However, despite a number of studies addressing this issue, there is no consensus on the role of EUS in routine clinical practice. Objectives To provide both a comprehensive overview and a quantitative analysis of the published data regarding the ability of EUS to preoperatively define the locoregional disease spread (i.e., primary tumor depth (T‐stage) and regional lymph node status (N‐stage)) in people with primary gastric carcinoma. Search methods We performed a systematic search to identify articles that examined the diagnostic accuracy of EUS (the index test) in the evaluation of primary gastric cancer depth of invasion (T‐stage, according to the AJCC/UICC TNM staging system categories T1, T2, T3 and T4) and regional lymph node status (N‐stage, disease‐free (N0) versus metastatic (N+)) using histopathology as the reference standard. To this end, we searched the following databases: the Cochrane Library (the Cochrane Central Register of Controlled Trials (CENTRAL)), MEDLINE, EMBASE , NIHR Prospero Register, MEDION, Aggressive Research Intelligence Facility (ARIF), ClinicalTrials.gov, Current Controlled Trials MetaRegister, and World Health Organization International Clinical Trials Registry Platform (WHO ICTRP), from 1988 to January 2015. Selection criteria We included studies that met the following main inclusion criteria: 1) a minimum sample size of 10 patients with histologically‐proven primary carcinoma of the stomach (target condition); 2) comparison of EUS (index test) with pathology evaluation (reference standard) in terms of primary tumor (T‐stage) and regional lymph nodes (N‐stage). We excluded reports with possible overlap with the selected studies. Data collection and analysis For each study, two review authors extracted a standard set of data, using a dedicated data extraction form. We assessed data quality using a standard procedure according to the Quality Assessment of Diagnostic Accuracy Studies (QUADAS‐2) criteria. We performed diagnostic accuracy meta‐analysis using the hierarchical bivariate method. We identified 66 articles (published between 1988 and 2012) that were eligible according to the inclusion criteria. We collected the data on 7747 patients with gastric cancer who were staged with EUS. Overall the quality of the included studies was good: in particular, only five studies presented a high risk of index test interpretation bias and two studies presented a high risk of selection bias. For primary tumor (T) stage, results were stratified according to the depth of invasion of the gastric wall. The meta‐analysis of 50 studies (n = 4397) showed that the summary sensitivity and specificity of EUS in discriminating T1 to T2 (superficial) versus T3 to T4 (advanced) gastric carcinomas were 0.86 (95% confidence interval (CI) 0.81 to 0.90) and 0.90 (95% CI 0.87 to 0.93) respectively. For the diagnostic capacity of EUS to distinguish T1 (early gastric cancer, EGC) versus T2 (muscle‐infiltrating) tumors, the meta‐analysis of 46 studies (n = 2742) showed that the summary sensitivity and specificity were 0.85 (95% CI 0.78 to 0.91) and 0.90 (95% CI 0.85 to 0.93) respectively. When we addressed the capacity of EUS to distinguish between T1a (mucosal) versus T1b (submucosal) cancers the meta‐analysis of 20 studies (n = 3321) showed that the summary sensitivity and specificity were 0.87 (95% CI 0.81 to 0.92) and 0.75 (95% CI 0.62 to 0.84) respectively. Finally, for the metastatic involvement of lymph nodes (N‐stage), the meta‐analysis of 44 studies (n = 3573) showed that the summary sensitivity and specificity were 0.83 (95% CI 0.79 to 0.87) and 0.67 (95% CI 0.61 to 0.72), respectively. Overall, as demonstrated also by the Bayesian nomograms, which enable readers to calculate post‐test probabilities for any target condition prevalence, the EUS accuracy can be considered clinically useful to guide physicians in the locoregional staging of people with gastric cancer. However, it should be noted that between‐study heterogeneity was not negligible: unfortunately, we could not identify any consistent source of the observed heterogeneity. Therefore, all accuracy measures reported in the present work and summarizing the available evidence should be interpreted cautiously. Moreover, we must emphasize that the analysis of positive and negative likelihood values revealed that EUS diagnostic performance cannot be considered optimal either for disease confirmation or for exclusion, especially for the ability of EUS to distinguish T1a (mucosal) versus T1b (submucosal) cancers and positive versus negative lymph node status. By analyzing the data from the largest series ever considered, we found that the diagnostic accuracy of EUS might be considered clinically useful to guide physicians in the locoregional staging of people with gastric carcinoma. However, the heterogeneity of the results warrants special caution, as well as further investigation for the identification of factors influencing the outcome of this diagnostic tool. Moreover, physicians should be warned that EUS performance is lower in diagnosing superficial tumors (T1a versus T1b) and lymph node status (positive versus negative). Overall, we observed large heterogeneity and its source needs to be understood before any definitive conclusion can be drawn about the use of EUS can be proposed in routine clinical settings.
t143	t143_8	no	We included 66 studies (of 7747 patients) in the review.	Endoscopic ultrasound (EUS) is proposed as an accurate diagnostic device for the locoregional staging of gastric cancer, which is crucial to developing a correct therapeutic strategy and ultimately to providing patients with the best chance of cure. However, despite a number of studies addressing this issue, there is no consensus on the role of EUS in routine clinical practice. Objectives To provide both a comprehensive overview and a quantitative analysis of the published data regarding the ability of EUS to preoperatively define the locoregional disease spread (i.e., primary tumor depth (T‐stage) and regional lymph node status (N‐stage)) in people with primary gastric carcinoma. Search methods We performed a systematic search to identify articles that examined the diagnostic accuracy of EUS (the index test) in the evaluation of primary gastric cancer depth of invasion (T‐stage, according to the AJCC/UICC TNM staging system categories T1, T2, T3 and T4) and regional lymph node status (N‐stage, disease‐free (N0) versus metastatic (N+)) using histopathology as the reference standard. To this end, we searched the following databases: the Cochrane Library (the Cochrane Central Register of Controlled Trials (CENTRAL)), MEDLINE, EMBASE , NIHR Prospero Register, MEDION, Aggressive Research Intelligence Facility (ARIF), ClinicalTrials.gov, Current Controlled Trials MetaRegister, and World Health Organization International Clinical Trials Registry Platform (WHO ICTRP), from 1988 to January 2015. Selection criteria We included studies that met the following main inclusion criteria: 1) a minimum sample size of 10 patients with histologically‐proven primary carcinoma of the stomach (target condition); 2) comparison of EUS (index test) with pathology evaluation (reference standard) in terms of primary tumor (T‐stage) and regional lymph nodes (N‐stage). We excluded reports with possible overlap with the selected studies. Data collection and analysis For each study, two review authors extracted a standard set of data, using a dedicated data extraction form. We assessed data quality using a standard procedure according to the Quality Assessment of Diagnostic Accuracy Studies (QUADAS‐2) criteria. We performed diagnostic accuracy meta‐analysis using the hierarchical bivariate method. We identified 66 articles (published between 1988 and 2012) that were eligible according to the inclusion criteria. We collected the data on 7747 patients with gastric cancer who were staged with EUS. Overall the quality of the included studies was good: in particular, only five studies presented a high risk of index test interpretation bias and two studies presented a high risk of selection bias. For primary tumor (T) stage, results were stratified according to the depth of invasion of the gastric wall. The meta‐analysis of 50 studies (n = 4397) showed that the summary sensitivity and specificity of EUS in discriminating T1 to T2 (superficial) versus T3 to T4 (advanced) gastric carcinomas were 0.86 (95% confidence interval (CI) 0.81 to 0.90) and 0.90 (95% CI 0.87 to 0.93) respectively. For the diagnostic capacity of EUS to distinguish T1 (early gastric cancer, EGC) versus T2 (muscle‐infiltrating) tumors, the meta‐analysis of 46 studies (n = 2742) showed that the summary sensitivity and specificity were 0.85 (95% CI 0.78 to 0.91) and 0.90 (95% CI 0.85 to 0.93) respectively. When we addressed the capacity of EUS to distinguish between T1a (mucosal) versus T1b (submucosal) cancers the meta‐analysis of 20 studies (n = 3321) showed that the summary sensitivity and specificity were 0.87 (95% CI 0.81 to 0.92) and 0.75 (95% CI 0.62 to 0.84) respectively. Finally, for the metastatic involvement of lymph nodes (N‐stage), the meta‐analysis of 44 studies (n = 3573) showed that the summary sensitivity and specificity were 0.83 (95% CI 0.79 to 0.87) and 0.67 (95% CI 0.61 to 0.72), respectively. Overall, as demonstrated also by the Bayesian nomograms, which enable readers to calculate post‐test probabilities for any target condition prevalence, the EUS accuracy can be considered clinically useful to guide physicians in the locoregional staging of people with gastric cancer. However, it should be noted that between‐study heterogeneity was not negligible: unfortunately, we could not identify any consistent source of the observed heterogeneity. Therefore, all accuracy measures reported in the present work and summarizing the available evidence should be interpreted cautiously. Moreover, we must emphasize that the analysis of positive and negative likelihood values revealed that EUS diagnostic performance cannot be considered optimal either for disease confirmation or for exclusion, especially for the ability of EUS to distinguish T1a (mucosal) versus T1b (submucosal) cancers and positive versus negative lymph node status. By analyzing the data from the largest series ever considered, we found that the diagnostic accuracy of EUS might be considered clinically useful to guide physicians in the locoregional staging of people with gastric carcinoma. However, the heterogeneity of the results warrants special caution, as well as further investigation for the identification of factors influencing the outcome of this diagnostic tool. Moreover, physicians should be warned that EUS performance is lower in diagnosing superficial tumors (T1a versus T1b) and lymph node status (positive versus negative). Overall, we observed large heterogeneity and its source needs to be understood before any definitive conclusion can be drawn about the use of EUS can be proposed in routine clinical settings.
t143	t143_9	no	We found that EUS can distinguish between superficial (T1 ‐ T2) and advanced (T3 ‐ T4) primary tumors with a sensitivity and a specificity greater than 85%.	Endoscopic ultrasound (EUS) is proposed as an accurate diagnostic device for the locoregional staging of gastric cancer, which is crucial to developing a correct therapeutic strategy and ultimately to providing patients with the best chance of cure. However, despite a number of studies addressing this issue, there is no consensus on the role of EUS in routine clinical practice. Objectives To provide both a comprehensive overview and a quantitative analysis of the published data regarding the ability of EUS to preoperatively define the locoregional disease spread (i.e., primary tumor depth (T‐stage) and regional lymph node status (N‐stage)) in people with primary gastric carcinoma. Search methods We performed a systematic search to identify articles that examined the diagnostic accuracy of EUS (the index test) in the evaluation of primary gastric cancer depth of invasion (T‐stage, according to the AJCC/UICC TNM staging system categories T1, T2, T3 and T4) and regional lymph node status (N‐stage, disease‐free (N0) versus metastatic (N+)) using histopathology as the reference standard. To this end, we searched the following databases: the Cochrane Library (the Cochrane Central Register of Controlled Trials (CENTRAL)), MEDLINE, EMBASE , NIHR Prospero Register, MEDION, Aggressive Research Intelligence Facility (ARIF), ClinicalTrials.gov, Current Controlled Trials MetaRegister, and World Health Organization International Clinical Trials Registry Platform (WHO ICTRP), from 1988 to January 2015. Selection criteria We included studies that met the following main inclusion criteria: 1) a minimum sample size of 10 patients with histologically‐proven primary carcinoma of the stomach (target condition); 2) comparison of EUS (index test) with pathology evaluation (reference standard) in terms of primary tumor (T‐stage) and regional lymph nodes (N‐stage). We excluded reports with possible overlap with the selected studies. Data collection and analysis For each study, two review authors extracted a standard set of data, using a dedicated data extraction form. We assessed data quality using a standard procedure according to the Quality Assessment of Diagnostic Accuracy Studies (QUADAS‐2) criteria. We performed diagnostic accuracy meta‐analysis using the hierarchical bivariate method. We identified 66 articles (published between 1988 and 2012) that were eligible according to the inclusion criteria. We collected the data on 7747 patients with gastric cancer who were staged with EUS. Overall the quality of the included studies was good: in particular, only five studies presented a high risk of index test interpretation bias and two studies presented a high risk of selection bias. For primary tumor (T) stage, results were stratified according to the depth of invasion of the gastric wall. The meta‐analysis of 50 studies (n = 4397) showed that the summary sensitivity and specificity of EUS in discriminating T1 to T2 (superficial) versus T3 to T4 (advanced) gastric carcinomas were 0.86 (95% confidence interval (CI) 0.81 to 0.90) and 0.90 (95% CI 0.87 to 0.93) respectively. For the diagnostic capacity of EUS to distinguish T1 (early gastric cancer, EGC) versus T2 (muscle‐infiltrating) tumors, the meta‐analysis of 46 studies (n = 2742) showed that the summary sensitivity and specificity were 0.85 (95% CI 0.78 to 0.91) and 0.90 (95% CI 0.85 to 0.93) respectively. When we addressed the capacity of EUS to distinguish between T1a (mucosal) versus T1b (submucosal) cancers the meta‐analysis of 20 studies (n = 3321) showed that the summary sensitivity and specificity were 0.87 (95% CI 0.81 to 0.92) and 0.75 (95% CI 0.62 to 0.84) respectively. Finally, for the metastatic involvement of lymph nodes (N‐stage), the meta‐analysis of 44 studies (n = 3573) showed that the summary sensitivity and specificity were 0.83 (95% CI 0.79 to 0.87) and 0.67 (95% CI 0.61 to 0.72), respectively. Overall, as demonstrated also by the Bayesian nomograms, which enable readers to calculate post‐test probabilities for any target condition prevalence, the EUS accuracy can be considered clinically useful to guide physicians in the locoregional staging of people with gastric cancer. However, it should be noted that between‐study heterogeneity was not negligible: unfortunately, we could not identify any consistent source of the observed heterogeneity. Therefore, all accuracy measures reported in the present work and summarizing the available evidence should be interpreted cautiously. Moreover, we must emphasize that the analysis of positive and negative likelihood values revealed that EUS diagnostic performance cannot be considered optimal either for disease confirmation or for exclusion, especially for the ability of EUS to distinguish T1a (mucosal) versus T1b (submucosal) cancers and positive versus negative lymph node status. By analyzing the data from the largest series ever considered, we found that the diagnostic accuracy of EUS might be considered clinically useful to guide physicians in the locoregional staging of people with gastric carcinoma. However, the heterogeneity of the results warrants special caution, as well as further investigation for the identification of factors influencing the outcome of this diagnostic tool. Moreover, physicians should be warned that EUS performance is lower in diagnosing superficial tumors (T1a versus T1b) and lymph node status (positive versus negative). Overall, we observed large heterogeneity and its source needs to be understood before any definitive conclusion can be drawn about the use of EUS can be proposed in routine clinical settings.
t143	t143_10	no	This performance is maintained for the discrimination between T1 and T2 superficial tumors.	Endoscopic ultrasound (EUS) is proposed as an accurate diagnostic device for the locoregional staging of gastric cancer, which is crucial to developing a correct therapeutic strategy and ultimately to providing patients with the best chance of cure. However, despite a number of studies addressing this issue, there is no consensus on the role of EUS in routine clinical practice. Objectives To provide both a comprehensive overview and a quantitative analysis of the published data regarding the ability of EUS to preoperatively define the locoregional disease spread (i.e., primary tumor depth (T‐stage) and regional lymph node status (N‐stage)) in people with primary gastric carcinoma. Search methods We performed a systematic search to identify articles that examined the diagnostic accuracy of EUS (the index test) in the evaluation of primary gastric cancer depth of invasion (T‐stage, according to the AJCC/UICC TNM staging system categories T1, T2, T3 and T4) and regional lymph node status (N‐stage, disease‐free (N0) versus metastatic (N+)) using histopathology as the reference standard. To this end, we searched the following databases: the Cochrane Library (the Cochrane Central Register of Controlled Trials (CENTRAL)), MEDLINE, EMBASE , NIHR Prospero Register, MEDION, Aggressive Research Intelligence Facility (ARIF), ClinicalTrials.gov, Current Controlled Trials MetaRegister, and World Health Organization International Clinical Trials Registry Platform (WHO ICTRP), from 1988 to January 2015. Selection criteria We included studies that met the following main inclusion criteria: 1) a minimum sample size of 10 patients with histologically‐proven primary carcinoma of the stomach (target condition); 2) comparison of EUS (index test) with pathology evaluation (reference standard) in terms of primary tumor (T‐stage) and regional lymph nodes (N‐stage). We excluded reports with possible overlap with the selected studies. Data collection and analysis For each study, two review authors extracted a standard set of data, using a dedicated data extraction form. We assessed data quality using a standard procedure according to the Quality Assessment of Diagnostic Accuracy Studies (QUADAS‐2) criteria. We performed diagnostic accuracy meta‐analysis using the hierarchical bivariate method. We identified 66 articles (published between 1988 and 2012) that were eligible according to the inclusion criteria. We collected the data on 7747 patients with gastric cancer who were staged with EUS. Overall the quality of the included studies was good: in particular, only five studies presented a high risk of index test interpretation bias and two studies presented a high risk of selection bias. For primary tumor (T) stage, results were stratified according to the depth of invasion of the gastric wall. The meta‐analysis of 50 studies (n = 4397) showed that the summary sensitivity and specificity of EUS in discriminating T1 to T2 (superficial) versus T3 to T4 (advanced) gastric carcinomas were 0.86 (95% confidence interval (CI) 0.81 to 0.90) and 0.90 (95% CI 0.87 to 0.93) respectively. For the diagnostic capacity of EUS to distinguish T1 (early gastric cancer, EGC) versus T2 (muscle‐infiltrating) tumors, the meta‐analysis of 46 studies (n = 2742) showed that the summary sensitivity and specificity were 0.85 (95% CI 0.78 to 0.91) and 0.90 (95% CI 0.85 to 0.93) respectively. When we addressed the capacity of EUS to distinguish between T1a (mucosal) versus T1b (submucosal) cancers the meta‐analysis of 20 studies (n = 3321) showed that the summary sensitivity and specificity were 0.87 (95% CI 0.81 to 0.92) and 0.75 (95% CI 0.62 to 0.84) respectively. Finally, for the metastatic involvement of lymph nodes (N‐stage), the meta‐analysis of 44 studies (n = 3573) showed that the summary sensitivity and specificity were 0.83 (95% CI 0.79 to 0.87) and 0.67 (95% CI 0.61 to 0.72), respectively. Overall, as demonstrated also by the Bayesian nomograms, which enable readers to calculate post‐test probabilities for any target condition prevalence, the EUS accuracy can be considered clinically useful to guide physicians in the locoregional staging of people with gastric cancer. However, it should be noted that between‐study heterogeneity was not negligible: unfortunately, we could not identify any consistent source of the observed heterogeneity. Therefore, all accuracy measures reported in the present work and summarizing the available evidence should be interpreted cautiously. Moreover, we must emphasize that the analysis of positive and negative likelihood values revealed that EUS diagnostic performance cannot be considered optimal either for disease confirmation or for exclusion, especially for the ability of EUS to distinguish T1a (mucosal) versus T1b (submucosal) cancers and positive versus negative lymph node status. By analyzing the data from the largest series ever considered, we found that the diagnostic accuracy of EUS might be considered clinically useful to guide physicians in the locoregional staging of people with gastric carcinoma. However, the heterogeneity of the results warrants special caution, as well as further investigation for the identification of factors influencing the outcome of this diagnostic tool. Moreover, physicians should be warned that EUS performance is lower in diagnosing superficial tumors (T1a versus T1b) and lymph node status (positive versus negative). Overall, we observed large heterogeneity and its source needs to be understood before any definitive conclusion can be drawn about the use of EUS can be proposed in routine clinical settings.
t143	t143_11	no	However, EUS diagnostic accuracy is lower when it comes to distinguishing between the different types of early tumors (T1a versus T1b) and between tumors with versus those without lymph node disease.	Endoscopic ultrasound (EUS) is proposed as an accurate diagnostic device for the locoregional staging of gastric cancer, which is crucial to developing a correct therapeutic strategy and ultimately to providing patients with the best chance of cure. However, despite a number of studies addressing this issue, there is no consensus on the role of EUS in routine clinical practice. Objectives To provide both a comprehensive overview and a quantitative analysis of the published data regarding the ability of EUS to preoperatively define the locoregional disease spread (i.e., primary tumor depth (T‐stage) and regional lymph node status (N‐stage)) in people with primary gastric carcinoma. Search methods We performed a systematic search to identify articles that examined the diagnostic accuracy of EUS (the index test) in the evaluation of primary gastric cancer depth of invasion (T‐stage, according to the AJCC/UICC TNM staging system categories T1, T2, T3 and T4) and regional lymph node status (N‐stage, disease‐free (N0) versus metastatic (N+)) using histopathology as the reference standard. To this end, we searched the following databases: the Cochrane Library (the Cochrane Central Register of Controlled Trials (CENTRAL)), MEDLINE, EMBASE , NIHR Prospero Register, MEDION, Aggressive Research Intelligence Facility (ARIF), ClinicalTrials.gov, Current Controlled Trials MetaRegister, and World Health Organization International Clinical Trials Registry Platform (WHO ICTRP), from 1988 to January 2015. Selection criteria We included studies that met the following main inclusion criteria: 1) a minimum sample size of 10 patients with histologically‐proven primary carcinoma of the stomach (target condition); 2) comparison of EUS (index test) with pathology evaluation (reference standard) in terms of primary tumor (T‐stage) and regional lymph nodes (N‐stage). We excluded reports with possible overlap with the selected studies. Data collection and analysis For each study, two review authors extracted a standard set of data, using a dedicated data extraction form. We assessed data quality using a standard procedure according to the Quality Assessment of Diagnostic Accuracy Studies (QUADAS‐2) criteria. We performed diagnostic accuracy meta‐analysis using the hierarchical bivariate method. We identified 66 articles (published between 1988 and 2012) that were eligible according to the inclusion criteria. We collected the data on 7747 patients with gastric cancer who were staged with EUS. Overall the quality of the included studies was good: in particular, only five studies presented a high risk of index test interpretation bias and two studies presented a high risk of selection bias. For primary tumor (T) stage, results were stratified according to the depth of invasion of the gastric wall. The meta‐analysis of 50 studies (n = 4397) showed that the summary sensitivity and specificity of EUS in discriminating T1 to T2 (superficial) versus T3 to T4 (advanced) gastric carcinomas were 0.86 (95% confidence interval (CI) 0.81 to 0.90) and 0.90 (95% CI 0.87 to 0.93) respectively. For the diagnostic capacity of EUS to distinguish T1 (early gastric cancer, EGC) versus T2 (muscle‐infiltrating) tumors, the meta‐analysis of 46 studies (n = 2742) showed that the summary sensitivity and specificity were 0.85 (95% CI 0.78 to 0.91) and 0.90 (95% CI 0.85 to 0.93) respectively. When we addressed the capacity of EUS to distinguish between T1a (mucosal) versus T1b (submucosal) cancers the meta‐analysis of 20 studies (n = 3321) showed that the summary sensitivity and specificity were 0.87 (95% CI 0.81 to 0.92) and 0.75 (95% CI 0.62 to 0.84) respectively. Finally, for the metastatic involvement of lymph nodes (N‐stage), the meta‐analysis of 44 studies (n = 3573) showed that the summary sensitivity and specificity were 0.83 (95% CI 0.79 to 0.87) and 0.67 (95% CI 0.61 to 0.72), respectively. Overall, as demonstrated also by the Bayesian nomograms, which enable readers to calculate post‐test probabilities for any target condition prevalence, the EUS accuracy can be considered clinically useful to guide physicians in the locoregional staging of people with gastric cancer. However, it should be noted that between‐study heterogeneity was not negligible: unfortunately, we could not identify any consistent source of the observed heterogeneity. Therefore, all accuracy measures reported in the present work and summarizing the available evidence should be interpreted cautiously. Moreover, we must emphasize that the analysis of positive and negative likelihood values revealed that EUS diagnostic performance cannot be considered optimal either for disease confirmation or for exclusion, especially for the ability of EUS to distinguish T1a (mucosal) versus T1b (submucosal) cancers and positive versus negative lymph node status. By analyzing the data from the largest series ever considered, we found that the diagnostic accuracy of EUS might be considered clinically useful to guide physicians in the locoregional staging of people with gastric carcinoma. However, the heterogeneity of the results warrants special caution, as well as further investigation for the identification of factors influencing the outcome of this diagnostic tool. Moreover, physicians should be warned that EUS performance is lower in diagnosing superficial tumors (T1a versus T1b) and lymph node status (positive versus negative). Overall, we observed large heterogeneity and its source needs to be understood before any definitive conclusion can be drawn about the use of EUS can be proposed in routine clinical settings.
t143	t143_12	no	Overall, EUS provides physicians with some helpful information on the stage of gastric cancer.	Endoscopic ultrasound (EUS) is proposed as an accurate diagnostic device for the locoregional staging of gastric cancer, which is crucial to developing a correct therapeutic strategy and ultimately to providing patients with the best chance of cure. However, despite a number of studies addressing this issue, there is no consensus on the role of EUS in routine clinical practice. Objectives To provide both a comprehensive overview and a quantitative analysis of the published data regarding the ability of EUS to preoperatively define the locoregional disease spread (i.e., primary tumor depth (T‐stage) and regional lymph node status (N‐stage)) in people with primary gastric carcinoma. Search methods We performed a systematic search to identify articles that examined the diagnostic accuracy of EUS (the index test) in the evaluation of primary gastric cancer depth of invasion (T‐stage, according to the AJCC/UICC TNM staging system categories T1, T2, T3 and T4) and regional lymph node status (N‐stage, disease‐free (N0) versus metastatic (N+)) using histopathology as the reference standard. To this end, we searched the following databases: the Cochrane Library (the Cochrane Central Register of Controlled Trials (CENTRAL)), MEDLINE, EMBASE , NIHR Prospero Register, MEDION, Aggressive Research Intelligence Facility (ARIF), ClinicalTrials.gov, Current Controlled Trials MetaRegister, and World Health Organization International Clinical Trials Registry Platform (WHO ICTRP), from 1988 to January 2015. Selection criteria We included studies that met the following main inclusion criteria: 1) a minimum sample size of 10 patients with histologically‐proven primary carcinoma of the stomach (target condition); 2) comparison of EUS (index test) with pathology evaluation (reference standard) in terms of primary tumor (T‐stage) and regional lymph nodes (N‐stage). We excluded reports with possible overlap with the selected studies. Data collection and analysis For each study, two review authors extracted a standard set of data, using a dedicated data extraction form. We assessed data quality using a standard procedure according to the Quality Assessment of Diagnostic Accuracy Studies (QUADAS‐2) criteria. We performed diagnostic accuracy meta‐analysis using the hierarchical bivariate method. We identified 66 articles (published between 1988 and 2012) that were eligible according to the inclusion criteria. We collected the data on 7747 patients with gastric cancer who were staged with EUS. Overall the quality of the included studies was good: in particular, only five studies presented a high risk of index test interpretation bias and two studies presented a high risk of selection bias. For primary tumor (T) stage, results were stratified according to the depth of invasion of the gastric wall. The meta‐analysis of 50 studies (n = 4397) showed that the summary sensitivity and specificity of EUS in discriminating T1 to T2 (superficial) versus T3 to T4 (advanced) gastric carcinomas were 0.86 (95% confidence interval (CI) 0.81 to 0.90) and 0.90 (95% CI 0.87 to 0.93) respectively. For the diagnostic capacity of EUS to distinguish T1 (early gastric cancer, EGC) versus T2 (muscle‐infiltrating) tumors, the meta‐analysis of 46 studies (n = 2742) showed that the summary sensitivity and specificity were 0.85 (95% CI 0.78 to 0.91) and 0.90 (95% CI 0.85 to 0.93) respectively. When we addressed the capacity of EUS to distinguish between T1a (mucosal) versus T1b (submucosal) cancers the meta‐analysis of 20 studies (n = 3321) showed that the summary sensitivity and specificity were 0.87 (95% CI 0.81 to 0.92) and 0.75 (95% CI 0.62 to 0.84) respectively. Finally, for the metastatic involvement of lymph nodes (N‐stage), the meta‐analysis of 44 studies (n = 3573) showed that the summary sensitivity and specificity were 0.83 (95% CI 0.79 to 0.87) and 0.67 (95% CI 0.61 to 0.72), respectively. Overall, as demonstrated also by the Bayesian nomograms, which enable readers to calculate post‐test probabilities for any target condition prevalence, the EUS accuracy can be considered clinically useful to guide physicians in the locoregional staging of people with gastric cancer. However, it should be noted that between‐study heterogeneity was not negligible: unfortunately, we could not identify any consistent source of the observed heterogeneity. Therefore, all accuracy measures reported in the present work and summarizing the available evidence should be interpreted cautiously. Moreover, we must emphasize that the analysis of positive and negative likelihood values revealed that EUS diagnostic performance cannot be considered optimal either for disease confirmation or for exclusion, especially for the ability of EUS to distinguish T1a (mucosal) versus T1b (submucosal) cancers and positive versus negative lymph node status. By analyzing the data from the largest series ever considered, we found that the diagnostic accuracy of EUS might be considered clinically useful to guide physicians in the locoregional staging of people with gastric carcinoma. However, the heterogeneity of the results warrants special caution, as well as further investigation for the identification of factors influencing the outcome of this diagnostic tool. Moreover, physicians should be warned that EUS performance is lower in diagnosing superficial tumors (T1a versus T1b) and lymph node status (positive versus negative). Overall, we observed large heterogeneity and its source needs to be understood before any definitive conclusion can be drawn about the use of EUS can be proposed in routine clinical settings.
t143	t143_13	no	Nevertheless, in the light of the variability of the results reported in the international medical literature, its limitations in terms of performance must be kept in mind in order to make the most out of the diagnostic potential of this tool.	Endoscopic ultrasound (EUS) is proposed as an accurate diagnostic device for the locoregional staging of gastric cancer, which is crucial to developing a correct therapeutic strategy and ultimately to providing patients with the best chance of cure. However, despite a number of studies addressing this issue, there is no consensus on the role of EUS in routine clinical practice. Objectives To provide both a comprehensive overview and a quantitative analysis of the published data regarding the ability of EUS to preoperatively define the locoregional disease spread (i.e., primary tumor depth (T‐stage) and regional lymph node status (N‐stage)) in people with primary gastric carcinoma. Search methods We performed a systematic search to identify articles that examined the diagnostic accuracy of EUS (the index test) in the evaluation of primary gastric cancer depth of invasion (T‐stage, according to the AJCC/UICC TNM staging system categories T1, T2, T3 and T4) and regional lymph node status (N‐stage, disease‐free (N0) versus metastatic (N+)) using histopathology as the reference standard. To this end, we searched the following databases: the Cochrane Library (the Cochrane Central Register of Controlled Trials (CENTRAL)), MEDLINE, EMBASE , NIHR Prospero Register, MEDION, Aggressive Research Intelligence Facility (ARIF), ClinicalTrials.gov, Current Controlled Trials MetaRegister, and World Health Organization International Clinical Trials Registry Platform (WHO ICTRP), from 1988 to January 2015. Selection criteria We included studies that met the following main inclusion criteria: 1) a minimum sample size of 10 patients with histologically‐proven primary carcinoma of the stomach (target condition); 2) comparison of EUS (index test) with pathology evaluation (reference standard) in terms of primary tumor (T‐stage) and regional lymph nodes (N‐stage). We excluded reports with possible overlap with the selected studies. Data collection and analysis For each study, two review authors extracted a standard set of data, using a dedicated data extraction form. We assessed data quality using a standard procedure according to the Quality Assessment of Diagnostic Accuracy Studies (QUADAS‐2) criteria. We performed diagnostic accuracy meta‐analysis using the hierarchical bivariate method. We identified 66 articles (published between 1988 and 2012) that were eligible according to the inclusion criteria. We collected the data on 7747 patients with gastric cancer who were staged with EUS. Overall the quality of the included studies was good: in particular, only five studies presented a high risk of index test interpretation bias and two studies presented a high risk of selection bias. For primary tumor (T) stage, results were stratified according to the depth of invasion of the gastric wall. The meta‐analysis of 50 studies (n = 4397) showed that the summary sensitivity and specificity of EUS in discriminating T1 to T2 (superficial) versus T3 to T4 (advanced) gastric carcinomas were 0.86 (95% confidence interval (CI) 0.81 to 0.90) and 0.90 (95% CI 0.87 to 0.93) respectively. For the diagnostic capacity of EUS to distinguish T1 (early gastric cancer, EGC) versus T2 (muscle‐infiltrating) tumors, the meta‐analysis of 46 studies (n = 2742) showed that the summary sensitivity and specificity were 0.85 (95% CI 0.78 to 0.91) and 0.90 (95% CI 0.85 to 0.93) respectively. When we addressed the capacity of EUS to distinguish between T1a (mucosal) versus T1b (submucosal) cancers the meta‐analysis of 20 studies (n = 3321) showed that the summary sensitivity and specificity were 0.87 (95% CI 0.81 to 0.92) and 0.75 (95% CI 0.62 to 0.84) respectively. Finally, for the metastatic involvement of lymph nodes (N‐stage), the meta‐analysis of 44 studies (n = 3573) showed that the summary sensitivity and specificity were 0.83 (95% CI 0.79 to 0.87) and 0.67 (95% CI 0.61 to 0.72), respectively. Overall, as demonstrated also by the Bayesian nomograms, which enable readers to calculate post‐test probabilities for any target condition prevalence, the EUS accuracy can be considered clinically useful to guide physicians in the locoregional staging of people with gastric cancer. However, it should be noted that between‐study heterogeneity was not negligible: unfortunately, we could not identify any consistent source of the observed heterogeneity. Therefore, all accuracy measures reported in the present work and summarizing the available evidence should be interpreted cautiously. Moreover, we must emphasize that the analysis of positive and negative likelihood values revealed that EUS diagnostic performance cannot be considered optimal either for disease confirmation or for exclusion, especially for the ability of EUS to distinguish T1a (mucosal) versus T1b (submucosal) cancers and positive versus negative lymph node status. By analyzing the data from the largest series ever considered, we found that the diagnostic accuracy of EUS might be considered clinically useful to guide physicians in the locoregional staging of people with gastric carcinoma. However, the heterogeneity of the results warrants special caution, as well as further investigation for the identification of factors influencing the outcome of this diagnostic tool. Moreover, physicians should be warned that EUS performance is lower in diagnosing superficial tumors (T1a versus T1b) and lymph node status (positive versus negative). Overall, we observed large heterogeneity and its source needs to be understood before any definitive conclusion can be drawn about the use of EUS can be proposed in routine clinical settings.
t144	t144_1	no	To find out what strategies can be used to improve how well healthcare workers follow a system of actions known as 'Standard Precautions' to decrease infection in healthcare settings.	'Standard Precautions' refers to a system of actions, such as using personal protective equipment or adhering to safe handling of needles, that healthcare workers take to reduce the spread of germs in healthcare settings such as hospitals and nursing homes. Objectives To assess the effectiveness of interventions that target healthcare workers to improve adherence to Standard Precautions in patient care. Search methods We searched CENTRAL, MEDLINE, Embase, CINAHL, LILACS, two other databases, and two trials registers. We applied no language restrictions. The date of the most recent search was 14 February 2017. Selection criteria We included randomised trials of individuals, cluster‐randomised trials, non‐randomised trials, controlled before‐after studies, and interrupted time‐series studies that evaluated any intervention to improve adherence to Standard Precautions by any healthcare worker with responsibility for patient care in any hospital, long‐term care or community setting, or artificial setting, such as a classroom or a learning laboratory. Data collection and analysis Two review authors independently screened search results, extracted data from eligible trials, and assessed risk of bias for each included study, using standard methodological procedures expected by Cochrane. Because of substantial heterogeneity among interventions and outcome measures, meta‐analysis was not warranted. We used the GRADE approach to assess certainty of evidence and have presented results narratively in 'Summary of findings' tables. We included eight studies with a total of 673 participants; three studies were conducted in Asia, two in Europe, two in North America, and one in Australia. Five studies were randomised trials, two were cluster‐randomised trials, and one was a non‐randomised trial. Three studies compared different educational approaches versus no education, one study compared education with visualisation of respiratory particle dispersion versus education alone, two studies compared education with additional infection control support versus no intervention, one study compared peer evaluation versus no intervention, and one study evaluated use of a checklist and coloured cues. We considered all studies to be at high risk of bias with different risks. All eight studies used different measures to assess healthcare workers' adherence to Standard Precautions. Three studies also assessed healthcare workers' knowledge, and one measured rates of colonisation with methicillin‐resistant Staphylococcus aureus (MRSA) among residents and staff of long‐term care facilities. Because of heterogeneity in interventions and outcome measures, we did not conduct a meta‐analysis. Education may slightly improve both healthcare workers' adherence to Standard Precautions (three studies; four centres) and their level of knowledge (two studies; three centres; low certainty of evidence for both outcomes). Education with visualisation of respiratory particle dispersion probably improves healthcare workers' use of facial protection but probably leads to little or no difference in knowledge (one study; 20 nurses; moderate certainty of evidence for both outcomes). Education with additional infection control support may slightly improve healthcare workers' adherence to Standard Precautions (two studies; 44 long‐term care facilities; low certainty of evidence) but probably leads to little or no difference in rates of health care‐associated colonisation with MRSA (one study; 32 long‐term care facilities; moderate certainty of evidence). Peer evaluation probably improves healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Checklists and coloured cues probably improve healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Considerable variation in interventions and in outcome measures used, along with high risk of bias and variability in the certainty of evidence, makes it difficult to draw conclusions about effectiveness of the interventions. This review underlines the need to conduct more robust studies evaluating similar types of interventions and using similar outcome measures.
t144	t144_2	no	Review authors identified a variety of strategies, most of which involved education of healthcare workers alone or with an additional strategy.	'Standard Precautions' refers to a system of actions, such as using personal protective equipment or adhering to safe handling of needles, that healthcare workers take to reduce the spread of germs in healthcare settings such as hospitals and nursing homes. Objectives To assess the effectiveness of interventions that target healthcare workers to improve adherence to Standard Precautions in patient care. Search methods We searched CENTRAL, MEDLINE, Embase, CINAHL, LILACS, two other databases, and two trials registers. We applied no language restrictions. The date of the most recent search was 14 February 2017. Selection criteria We included randomised trials of individuals, cluster‐randomised trials, non‐randomised trials, controlled before‐after studies, and interrupted time‐series studies that evaluated any intervention to improve adherence to Standard Precautions by any healthcare worker with responsibility for patient care in any hospital, long‐term care or community setting, or artificial setting, such as a classroom or a learning laboratory. Data collection and analysis Two review authors independently screened search results, extracted data from eligible trials, and assessed risk of bias for each included study, using standard methodological procedures expected by Cochrane. Because of substantial heterogeneity among interventions and outcome measures, meta‐analysis was not warranted. We used the GRADE approach to assess certainty of evidence and have presented results narratively in 'Summary of findings' tables. We included eight studies with a total of 673 participants; three studies were conducted in Asia, two in Europe, two in North America, and one in Australia. Five studies were randomised trials, two were cluster‐randomised trials, and one was a non‐randomised trial. Three studies compared different educational approaches versus no education, one study compared education with visualisation of respiratory particle dispersion versus education alone, two studies compared education with additional infection control support versus no intervention, one study compared peer evaluation versus no intervention, and one study evaluated use of a checklist and coloured cues. We considered all studies to be at high risk of bias with different risks. All eight studies used different measures to assess healthcare workers' adherence to Standard Precautions. Three studies also assessed healthcare workers' knowledge, and one measured rates of colonisation with methicillin‐resistant Staphylococcus aureus (MRSA) among residents and staff of long‐term care facilities. Because of heterogeneity in interventions and outcome measures, we did not conduct a meta‐analysis. Education may slightly improve both healthcare workers' adherence to Standard Precautions (three studies; four centres) and their level of knowledge (two studies; three centres; low certainty of evidence for both outcomes). Education with visualisation of respiratory particle dispersion probably improves healthcare workers' use of facial protection but probably leads to little or no difference in knowledge (one study; 20 nurses; moderate certainty of evidence for both outcomes). Education with additional infection control support may slightly improve healthcare workers' adherence to Standard Precautions (two studies; 44 long‐term care facilities; low certainty of evidence) but probably leads to little or no difference in rates of health care‐associated colonisation with MRSA (one study; 32 long‐term care facilities; moderate certainty of evidence). Peer evaluation probably improves healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Checklists and coloured cues probably improve healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Considerable variation in interventions and in outcome measures used, along with high risk of bias and variability in the certainty of evidence, makes it difficult to draw conclusions about effectiveness of the interventions. This review underlines the need to conduct more robust studies evaluating similar types of interventions and using similar outcome measures.
t144	t144_3	no	It is unclear which strategy or combination of strategies is most effective for improving healthcare workers' adherence to Standard Precautions or their knowledge of Standard Precautions, or for reducing colonisation (potential infection) rates, as we found little evidence; this fact, along with the inconsistency of results, reduced our confidence or certainty about the evidence found.	'Standard Precautions' refers to a system of actions, such as using personal protective equipment or adhering to safe handling of needles, that healthcare workers take to reduce the spread of germs in healthcare settings such as hospitals and nursing homes. Objectives To assess the effectiveness of interventions that target healthcare workers to improve adherence to Standard Precautions in patient care. Search methods We searched CENTRAL, MEDLINE, Embase, CINAHL, LILACS, two other databases, and two trials registers. We applied no language restrictions. The date of the most recent search was 14 February 2017. Selection criteria We included randomised trials of individuals, cluster‐randomised trials, non‐randomised trials, controlled before‐after studies, and interrupted time‐series studies that evaluated any intervention to improve adherence to Standard Precautions by any healthcare worker with responsibility for patient care in any hospital, long‐term care or community setting, or artificial setting, such as a classroom or a learning laboratory. Data collection and analysis Two review authors independently screened search results, extracted data from eligible trials, and assessed risk of bias for each included study, using standard methodological procedures expected by Cochrane. Because of substantial heterogeneity among interventions and outcome measures, meta‐analysis was not warranted. We used the GRADE approach to assess certainty of evidence and have presented results narratively in 'Summary of findings' tables. We included eight studies with a total of 673 participants; three studies were conducted in Asia, two in Europe, two in North America, and one in Australia. Five studies were randomised trials, two were cluster‐randomised trials, and one was a non‐randomised trial. Three studies compared different educational approaches versus no education, one study compared education with visualisation of respiratory particle dispersion versus education alone, two studies compared education with additional infection control support versus no intervention, one study compared peer evaluation versus no intervention, and one study evaluated use of a checklist and coloured cues. We considered all studies to be at high risk of bias with different risks. All eight studies used different measures to assess healthcare workers' adherence to Standard Precautions. Three studies also assessed healthcare workers' knowledge, and one measured rates of colonisation with methicillin‐resistant Staphylococcus aureus (MRSA) among residents and staff of long‐term care facilities. Because of heterogeneity in interventions and outcome measures, we did not conduct a meta‐analysis. Education may slightly improve both healthcare workers' adherence to Standard Precautions (three studies; four centres) and their level of knowledge (two studies; three centres; low certainty of evidence for both outcomes). Education with visualisation of respiratory particle dispersion probably improves healthcare workers' use of facial protection but probably leads to little or no difference in knowledge (one study; 20 nurses; moderate certainty of evidence for both outcomes). Education with additional infection control support may slightly improve healthcare workers' adherence to Standard Precautions (two studies; 44 long‐term care facilities; low certainty of evidence) but probably leads to little or no difference in rates of health care‐associated colonisation with MRSA (one study; 32 long‐term care facilities; moderate certainty of evidence). Peer evaluation probably improves healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Checklists and coloured cues probably improve healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Considerable variation in interventions and in outcome measures used, along with high risk of bias and variability in the certainty of evidence, makes it difficult to draw conclusions about effectiveness of the interventions. This review underlines the need to conduct more robust studies evaluating similar types of interventions and using similar outcome measures.
t144	t144_4	yes	It is estimated that over four million patients in Europe and 1.7 million in the USA develop an infection each year, and that prevalence is higher in developing countries.	'Standard Precautions' refers to a system of actions, such as using personal protective equipment or adhering to safe handling of needles, that healthcare workers take to reduce the spread of germs in healthcare settings such as hospitals and nursing homes. Objectives To assess the effectiveness of interventions that target healthcare workers to improve adherence to Standard Precautions in patient care. Search methods We searched CENTRAL, MEDLINE, Embase, CINAHL, LILACS, two other databases, and two trials registers. We applied no language restrictions. The date of the most recent search was 14 February 2017. Selection criteria We included randomised trials of individuals, cluster‐randomised trials, non‐randomised trials, controlled before‐after studies, and interrupted time‐series studies that evaluated any intervention to improve adherence to Standard Precautions by any healthcare worker with responsibility for patient care in any hospital, long‐term care or community setting, or artificial setting, such as a classroom or a learning laboratory. Data collection and analysis Two review authors independently screened search results, extracted data from eligible trials, and assessed risk of bias for each included study, using standard methodological procedures expected by Cochrane. Because of substantial heterogeneity among interventions and outcome measures, meta‐analysis was not warranted. We used the GRADE approach to assess certainty of evidence and have presented results narratively in 'Summary of findings' tables. We included eight studies with a total of 673 participants; three studies were conducted in Asia, two in Europe, two in North America, and one in Australia. Five studies were randomised trials, two were cluster‐randomised trials, and one was a non‐randomised trial. Three studies compared different educational approaches versus no education, one study compared education with visualisation of respiratory particle dispersion versus education alone, two studies compared education with additional infection control support versus no intervention, one study compared peer evaluation versus no intervention, and one study evaluated use of a checklist and coloured cues. We considered all studies to be at high risk of bias with different risks. All eight studies used different measures to assess healthcare workers' adherence to Standard Precautions. Three studies also assessed healthcare workers' knowledge, and one measured rates of colonisation with methicillin‐resistant Staphylococcus aureus (MRSA) among residents and staff of long‐term care facilities. Because of heterogeneity in interventions and outcome measures, we did not conduct a meta‐analysis. Education may slightly improve both healthcare workers' adherence to Standard Precautions (three studies; four centres) and their level of knowledge (two studies; three centres; low certainty of evidence for both outcomes). Education with visualisation of respiratory particle dispersion probably improves healthcare workers' use of facial protection but probably leads to little or no difference in knowledge (one study; 20 nurses; moderate certainty of evidence for both outcomes). Education with additional infection control support may slightly improve healthcare workers' adherence to Standard Precautions (two studies; 44 long‐term care facilities; low certainty of evidence) but probably leads to little or no difference in rates of health care‐associated colonisation with MRSA (one study; 32 long‐term care facilities; moderate certainty of evidence). Peer evaluation probably improves healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Checklists and coloured cues probably improve healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Considerable variation in interventions and in outcome measures used, along with high risk of bias and variability in the certainty of evidence, makes it difficult to draw conclusions about effectiveness of the interventions. This review underlines the need to conduct more robust studies evaluating similar types of interventions and using similar outcome measures.
t144	t144_5	yes	Infection is associated with increased length of hospital stay, excess mortality, and billions of dollars in associated hospital costs.	'Standard Precautions' refers to a system of actions, such as using personal protective equipment or adhering to safe handling of needles, that healthcare workers take to reduce the spread of germs in healthcare settings such as hospitals and nursing homes. Objectives To assess the effectiveness of interventions that target healthcare workers to improve adherence to Standard Precautions in patient care. Search methods We searched CENTRAL, MEDLINE, Embase, CINAHL, LILACS, two other databases, and two trials registers. We applied no language restrictions. The date of the most recent search was 14 February 2017. Selection criteria We included randomised trials of individuals, cluster‐randomised trials, non‐randomised trials, controlled before‐after studies, and interrupted time‐series studies that evaluated any intervention to improve adherence to Standard Precautions by any healthcare worker with responsibility for patient care in any hospital, long‐term care or community setting, or artificial setting, such as a classroom or a learning laboratory. Data collection and analysis Two review authors independently screened search results, extracted data from eligible trials, and assessed risk of bias for each included study, using standard methodological procedures expected by Cochrane. Because of substantial heterogeneity among interventions and outcome measures, meta‐analysis was not warranted. We used the GRADE approach to assess certainty of evidence and have presented results narratively in 'Summary of findings' tables. We included eight studies with a total of 673 participants; three studies were conducted in Asia, two in Europe, two in North America, and one in Australia. Five studies were randomised trials, two were cluster‐randomised trials, and one was a non‐randomised trial. Three studies compared different educational approaches versus no education, one study compared education with visualisation of respiratory particle dispersion versus education alone, two studies compared education with additional infection control support versus no intervention, one study compared peer evaluation versus no intervention, and one study evaluated use of a checklist and coloured cues. We considered all studies to be at high risk of bias with different risks. All eight studies used different measures to assess healthcare workers' adherence to Standard Precautions. Three studies also assessed healthcare workers' knowledge, and one measured rates of colonisation with methicillin‐resistant Staphylococcus aureus (MRSA) among residents and staff of long‐term care facilities. Because of heterogeneity in interventions and outcome measures, we did not conduct a meta‐analysis. Education may slightly improve both healthcare workers' adherence to Standard Precautions (three studies; four centres) and their level of knowledge (two studies; three centres; low certainty of evidence for both outcomes). Education with visualisation of respiratory particle dispersion probably improves healthcare workers' use of facial protection but probably leads to little or no difference in knowledge (one study; 20 nurses; moderate certainty of evidence for both outcomes). Education with additional infection control support may slightly improve healthcare workers' adherence to Standard Precautions (two studies; 44 long‐term care facilities; low certainty of evidence) but probably leads to little or no difference in rates of health care‐associated colonisation with MRSA (one study; 32 long‐term care facilities; moderate certainty of evidence). Peer evaluation probably improves healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Checklists and coloured cues probably improve healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Considerable variation in interventions and in outcome measures used, along with high risk of bias and variability in the certainty of evidence, makes it difficult to draw conclusions about effectiveness of the interventions. This review underlines the need to conduct more robust studies evaluating similar types of interventions and using similar outcome measures.
t144	t144_6	no	Adhering to Standard Precautions, such as using personal protective equipment or following practices for safe handling of needles, can reduce the spread of germs in healthcare settings.	'Standard Precautions' refers to a system of actions, such as using personal protective equipment or adhering to safe handling of needles, that healthcare workers take to reduce the spread of germs in healthcare settings such as hospitals and nursing homes. Objectives To assess the effectiveness of interventions that target healthcare workers to improve adherence to Standard Precautions in patient care. Search methods We searched CENTRAL, MEDLINE, Embase, CINAHL, LILACS, two other databases, and two trials registers. We applied no language restrictions. The date of the most recent search was 14 February 2017. Selection criteria We included randomised trials of individuals, cluster‐randomised trials, non‐randomised trials, controlled before‐after studies, and interrupted time‐series studies that evaluated any intervention to improve adherence to Standard Precautions by any healthcare worker with responsibility for patient care in any hospital, long‐term care or community setting, or artificial setting, such as a classroom or a learning laboratory. Data collection and analysis Two review authors independently screened search results, extracted data from eligible trials, and assessed risk of bias for each included study, using standard methodological procedures expected by Cochrane. Because of substantial heterogeneity among interventions and outcome measures, meta‐analysis was not warranted. We used the GRADE approach to assess certainty of evidence and have presented results narratively in 'Summary of findings' tables. We included eight studies with a total of 673 participants; three studies were conducted in Asia, two in Europe, two in North America, and one in Australia. Five studies were randomised trials, two were cluster‐randomised trials, and one was a non‐randomised trial. Three studies compared different educational approaches versus no education, one study compared education with visualisation of respiratory particle dispersion versus education alone, two studies compared education with additional infection control support versus no intervention, one study compared peer evaluation versus no intervention, and one study evaluated use of a checklist and coloured cues. We considered all studies to be at high risk of bias with different risks. All eight studies used different measures to assess healthcare workers' adherence to Standard Precautions. Three studies also assessed healthcare workers' knowledge, and one measured rates of colonisation with methicillin‐resistant Staphylococcus aureus (MRSA) among residents and staff of long‐term care facilities. Because of heterogeneity in interventions and outcome measures, we did not conduct a meta‐analysis. Education may slightly improve both healthcare workers' adherence to Standard Precautions (three studies; four centres) and their level of knowledge (two studies; three centres; low certainty of evidence for both outcomes). Education with visualisation of respiratory particle dispersion probably improves healthcare workers' use of facial protection but probably leads to little or no difference in knowledge (one study; 20 nurses; moderate certainty of evidence for both outcomes). Education with additional infection control support may slightly improve healthcare workers' adherence to Standard Precautions (two studies; 44 long‐term care facilities; low certainty of evidence) but probably leads to little or no difference in rates of health care‐associated colonisation with MRSA (one study; 32 long‐term care facilities; moderate certainty of evidence). Peer evaluation probably improves healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Checklists and coloured cues probably improve healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Considerable variation in interventions and in outcome measures used, along with high risk of bias and variability in the certainty of evidence, makes it difficult to draw conclusions about effectiveness of the interventions. This review underlines the need to conduct more robust studies evaluating similar types of interventions and using similar outcome measures.
t144	t144_7	no	The aim of this review was to find out which methods are effective in improving healthcare workers' adherence to Standard Precautions.	'Standard Precautions' refers to a system of actions, such as using personal protective equipment or adhering to safe handling of needles, that healthcare workers take to reduce the spread of germs in healthcare settings such as hospitals and nursing homes. Objectives To assess the effectiveness of interventions that target healthcare workers to improve adherence to Standard Precautions in patient care. Search methods We searched CENTRAL, MEDLINE, Embase, CINAHL, LILACS, two other databases, and two trials registers. We applied no language restrictions. The date of the most recent search was 14 February 2017. Selection criteria We included randomised trials of individuals, cluster‐randomised trials, non‐randomised trials, controlled before‐after studies, and interrupted time‐series studies that evaluated any intervention to improve adherence to Standard Precautions by any healthcare worker with responsibility for patient care in any hospital, long‐term care or community setting, or artificial setting, such as a classroom or a learning laboratory. Data collection and analysis Two review authors independently screened search results, extracted data from eligible trials, and assessed risk of bias for each included study, using standard methodological procedures expected by Cochrane. Because of substantial heterogeneity among interventions and outcome measures, meta‐analysis was not warranted. We used the GRADE approach to assess certainty of evidence and have presented results narratively in 'Summary of findings' tables. We included eight studies with a total of 673 participants; three studies were conducted in Asia, two in Europe, two in North America, and one in Australia. Five studies were randomised trials, two were cluster‐randomised trials, and one was a non‐randomised trial. Three studies compared different educational approaches versus no education, one study compared education with visualisation of respiratory particle dispersion versus education alone, two studies compared education with additional infection control support versus no intervention, one study compared peer evaluation versus no intervention, and one study evaluated use of a checklist and coloured cues. We considered all studies to be at high risk of bias with different risks. All eight studies used different measures to assess healthcare workers' adherence to Standard Precautions. Three studies also assessed healthcare workers' knowledge, and one measured rates of colonisation with methicillin‐resistant Staphylococcus aureus (MRSA) among residents and staff of long‐term care facilities. Because of heterogeneity in interventions and outcome measures, we did not conduct a meta‐analysis. Education may slightly improve both healthcare workers' adherence to Standard Precautions (three studies; four centres) and their level of knowledge (two studies; three centres; low certainty of evidence for both outcomes). Education with visualisation of respiratory particle dispersion probably improves healthcare workers' use of facial protection but probably leads to little or no difference in knowledge (one study; 20 nurses; moderate certainty of evidence for both outcomes). Education with additional infection control support may slightly improve healthcare workers' adherence to Standard Precautions (two studies; 44 long‐term care facilities; low certainty of evidence) but probably leads to little or no difference in rates of health care‐associated colonisation with MRSA (one study; 32 long‐term care facilities; moderate certainty of evidence). Peer evaluation probably improves healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Checklists and coloured cues probably improve healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Considerable variation in interventions and in outcome measures used, along with high risk of bias and variability in the certainty of evidence, makes it difficult to draw conclusions about effectiveness of the interventions. This review underlines the need to conduct more robust studies evaluating similar types of interventions and using similar outcome measures.
t144	t144_8	no	Review authors found eight relevant studies with a total of 673 participants.	'Standard Precautions' refers to a system of actions, such as using personal protective equipment or adhering to safe handling of needles, that healthcare workers take to reduce the spread of germs in healthcare settings such as hospitals and nursing homes. Objectives To assess the effectiveness of interventions that target healthcare workers to improve adherence to Standard Precautions in patient care. Search methods We searched CENTRAL, MEDLINE, Embase, CINAHL, LILACS, two other databases, and two trials registers. We applied no language restrictions. The date of the most recent search was 14 February 2017. Selection criteria We included randomised trials of individuals, cluster‐randomised trials, non‐randomised trials, controlled before‐after studies, and interrupted time‐series studies that evaluated any intervention to improve adherence to Standard Precautions by any healthcare worker with responsibility for patient care in any hospital, long‐term care or community setting, or artificial setting, such as a classroom or a learning laboratory. Data collection and analysis Two review authors independently screened search results, extracted data from eligible trials, and assessed risk of bias for each included study, using standard methodological procedures expected by Cochrane. Because of substantial heterogeneity among interventions and outcome measures, meta‐analysis was not warranted. We used the GRADE approach to assess certainty of evidence and have presented results narratively in 'Summary of findings' tables. We included eight studies with a total of 673 participants; three studies were conducted in Asia, two in Europe, two in North America, and one in Australia. Five studies were randomised trials, two were cluster‐randomised trials, and one was a non‐randomised trial. Three studies compared different educational approaches versus no education, one study compared education with visualisation of respiratory particle dispersion versus education alone, two studies compared education with additional infection control support versus no intervention, one study compared peer evaluation versus no intervention, and one study evaluated use of a checklist and coloured cues. We considered all studies to be at high risk of bias with different risks. All eight studies used different measures to assess healthcare workers' adherence to Standard Precautions. Three studies also assessed healthcare workers' knowledge, and one measured rates of colonisation with methicillin‐resistant Staphylococcus aureus (MRSA) among residents and staff of long‐term care facilities. Because of heterogeneity in interventions and outcome measures, we did not conduct a meta‐analysis. Education may slightly improve both healthcare workers' adherence to Standard Precautions (three studies; four centres) and their level of knowledge (two studies; three centres; low certainty of evidence for both outcomes). Education with visualisation of respiratory particle dispersion probably improves healthcare workers' use of facial protection but probably leads to little or no difference in knowledge (one study; 20 nurses; moderate certainty of evidence for both outcomes). Education with additional infection control support may slightly improve healthcare workers' adherence to Standard Precautions (two studies; 44 long‐term care facilities; low certainty of evidence) but probably leads to little or no difference in rates of health care‐associated colonisation with MRSA (one study; 32 long‐term care facilities; moderate certainty of evidence). Peer evaluation probably improves healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Checklists and coloured cues probably improve healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Considerable variation in interventions and in outcome measures used, along with high risk of bias and variability in the certainty of evidence, makes it difficult to draw conclusions about effectiveness of the interventions. This review underlines the need to conduct more robust studies evaluating similar types of interventions and using similar outcome measures.
t144	t144_9	no	Three studies were reported from Asia, two from Europe, two from North America, and one from Australia.	'Standard Precautions' refers to a system of actions, such as using personal protective equipment or adhering to safe handling of needles, that healthcare workers take to reduce the spread of germs in healthcare settings such as hospitals and nursing homes. Objectives To assess the effectiveness of interventions that target healthcare workers to improve adherence to Standard Precautions in patient care. Search methods We searched CENTRAL, MEDLINE, Embase, CINAHL, LILACS, two other databases, and two trials registers. We applied no language restrictions. The date of the most recent search was 14 February 2017. Selection criteria We included randomised trials of individuals, cluster‐randomised trials, non‐randomised trials, controlled before‐after studies, and interrupted time‐series studies that evaluated any intervention to improve adherence to Standard Precautions by any healthcare worker with responsibility for patient care in any hospital, long‐term care or community setting, or artificial setting, such as a classroom or a learning laboratory. Data collection and analysis Two review authors independently screened search results, extracted data from eligible trials, and assessed risk of bias for each included study, using standard methodological procedures expected by Cochrane. Because of substantial heterogeneity among interventions and outcome measures, meta‐analysis was not warranted. We used the GRADE approach to assess certainty of evidence and have presented results narratively in 'Summary of findings' tables. We included eight studies with a total of 673 participants; three studies were conducted in Asia, two in Europe, two in North America, and one in Australia. Five studies were randomised trials, two were cluster‐randomised trials, and one was a non‐randomised trial. Three studies compared different educational approaches versus no education, one study compared education with visualisation of respiratory particle dispersion versus education alone, two studies compared education with additional infection control support versus no intervention, one study compared peer evaluation versus no intervention, and one study evaluated use of a checklist and coloured cues. We considered all studies to be at high risk of bias with different risks. All eight studies used different measures to assess healthcare workers' adherence to Standard Precautions. Three studies also assessed healthcare workers' knowledge, and one measured rates of colonisation with methicillin‐resistant Staphylococcus aureus (MRSA) among residents and staff of long‐term care facilities. Because of heterogeneity in interventions and outcome measures, we did not conduct a meta‐analysis. Education may slightly improve both healthcare workers' adherence to Standard Precautions (three studies; four centres) and their level of knowledge (two studies; three centres; low certainty of evidence for both outcomes). Education with visualisation of respiratory particle dispersion probably improves healthcare workers' use of facial protection but probably leads to little or no difference in knowledge (one study; 20 nurses; moderate certainty of evidence for both outcomes). Education with additional infection control support may slightly improve healthcare workers' adherence to Standard Precautions (two studies; 44 long‐term care facilities; low certainty of evidence) but probably leads to little or no difference in rates of health care‐associated colonisation with MRSA (one study; 32 long‐term care facilities; moderate certainty of evidence). Peer evaluation probably improves healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Checklists and coloured cues probably improve healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Considerable variation in interventions and in outcome measures used, along with high risk of bias and variability in the certainty of evidence, makes it difficult to draw conclusions about effectiveness of the interventions. This review underlines the need to conduct more robust studies evaluating similar types of interventions and using similar outcome measures.
t144	t144_10	no	Intevention strategies consisted of education for healthcare workers, given alone or with other types of education, such as showing how respiratory droplets are spread, or with additional infection control supports.	'Standard Precautions' refers to a system of actions, such as using personal protective equipment or adhering to safe handling of needles, that healthcare workers take to reduce the spread of germs in healthcare settings such as hospitals and nursing homes. Objectives To assess the effectiveness of interventions that target healthcare workers to improve adherence to Standard Precautions in patient care. Search methods We searched CENTRAL, MEDLINE, Embase, CINAHL, LILACS, two other databases, and two trials registers. We applied no language restrictions. The date of the most recent search was 14 February 2017. Selection criteria We included randomised trials of individuals, cluster‐randomised trials, non‐randomised trials, controlled before‐after studies, and interrupted time‐series studies that evaluated any intervention to improve adherence to Standard Precautions by any healthcare worker with responsibility for patient care in any hospital, long‐term care or community setting, or artificial setting, such as a classroom or a learning laboratory. Data collection and analysis Two review authors independently screened search results, extracted data from eligible trials, and assessed risk of bias for each included study, using standard methodological procedures expected by Cochrane. Because of substantial heterogeneity among interventions and outcome measures, meta‐analysis was not warranted. We used the GRADE approach to assess certainty of evidence and have presented results narratively in 'Summary of findings' tables. We included eight studies with a total of 673 participants; three studies were conducted in Asia, two in Europe, two in North America, and one in Australia. Five studies were randomised trials, two were cluster‐randomised trials, and one was a non‐randomised trial. Three studies compared different educational approaches versus no education, one study compared education with visualisation of respiratory particle dispersion versus education alone, two studies compared education with additional infection control support versus no intervention, one study compared peer evaluation versus no intervention, and one study evaluated use of a checklist and coloured cues. We considered all studies to be at high risk of bias with different risks. All eight studies used different measures to assess healthcare workers' adherence to Standard Precautions. Three studies also assessed healthcare workers' knowledge, and one measured rates of colonisation with methicillin‐resistant Staphylococcus aureus (MRSA) among residents and staff of long‐term care facilities. Because of heterogeneity in interventions and outcome measures, we did not conduct a meta‐analysis. Education may slightly improve both healthcare workers' adherence to Standard Precautions (three studies; four centres) and their level of knowledge (two studies; three centres; low certainty of evidence for both outcomes). Education with visualisation of respiratory particle dispersion probably improves healthcare workers' use of facial protection but probably leads to little or no difference in knowledge (one study; 20 nurses; moderate certainty of evidence for both outcomes). Education with additional infection control support may slightly improve healthcare workers' adherence to Standard Precautions (two studies; 44 long‐term care facilities; low certainty of evidence) but probably leads to little or no difference in rates of health care‐associated colonisation with MRSA (one study; 32 long‐term care facilities; moderate certainty of evidence). Peer evaluation probably improves healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Checklists and coloured cues probably improve healthcare workers' adherence to Standard Precautions (one study; one hospital; moderate certainty of evidence). Considerable variation in interventions and in outcome measures used, along with high risk of bias and variability in the certainty of evidence, makes it difficult to draw conclusions about effectiveness of the interventions. This review underlines the need to conduct more robust studies evaluating similar types of interventions and using similar outcome measures.

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