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Intervention effect estimates in cluster randomized
versus individually randomized trials: a
Clémence Leyrat, Agnès Caille, Sandra Eldridge, Sally Kerry, Agnès
Dechartres, Bruno Giraudeau
To cite this version:
Clémence Leyrat, Agnès Caille, Sandra Eldridge, Sally Kerry, Agnès Dechartres, et al.. Intervention
effect estimates in cluster randomized versus individually randomized trials: a meta-epidemiological
study. International Journal of Epidemiology, Oxford University Press (OUP), 2019, 48 (2),
pp.609-619. �10.1093/ije/dyy229�. �hal-02282674�
Intervention effect estimates in cluster
randomized versus individually randomized
trials: a meta-epidemiological study
* Agne`s Caille,2,3
and Bruno Giraudeau2,3
Department of Medical Statistics, London School of Hygiene and Tropical Medicine, London, UK,2
Universite´ de Tours, Universite´ de Nantes, INSERM, SPHERE U1246, Tours, France,3
CHRU de Tours, Tours, France,4
Centre for Primary Care and Public Health, Queen Mary University of
London, London, UK and5
Sorbonne Universite´, INSERM, Institut Pierre Louis d’Epide´miologie et de
Sante´ Publique, De´partement Biostatistique Sante´ Publique et Information Me´dicale, Hoˆpitaux
Universitaires Pitie´ Salpeˆtrie`re – Charles Foix, AP-HP, Paris, France
*Corresponding author. Department of Medical Statistics, London School of Hygiene and Tropical Medicine, Keppel Street, WC1E 7HT London, UK. E-mail: email@example.com
Editorial decision 18 September 2018; Accepted 4 October 2018
Background: Cluster randomized trials (CRTs) and individually randomized trials (IRTs)
are often pooled together in meta-analyses (MAs) of randomized trials. However, the
po-tential systematic differences in intervention effect estimates between these two trial
types has never been investigated. Therefore, we conducted a meta-epidemiological
study comparing intervention effect estimates between CRTs and IRTs.
Methods: All Cochrane MAs including at least one CRT and one IRT, published between
1 January 2010 and 31 December 2014, were included. For each MA, we estimated a ratio
of odds ratios (ROR) for binary outcomes or a difference of standardized differences
(DSMD) for continuous outcomes, where less than 1 (or 0, respectively) indicated a
greater intervention effect estimate with CRTs.
Results: Among 1301 screened reviews, we selected 121 MAs, of which 76 had a binary
outcome and 45 had a continuous outcome. For binary outcomes, intervention effect
estimates did not differ between CRTs and IRTs [ROR 1.00, 95% confidence interval (0.93
to 1.08)]. Subgroup and adjusted analyses led to consistent results. For continuous
out-comes, the DSMD was 0.13 (0.06 to 0.19). It was lower for MAs with a pharmacological
intervention [-0.03, (-0.12 to 0.07)], an objective outcome [0.05, (-0.08 to 0.17)] or after
adjusting for trial size [0.06, (-0.01 to 0.15)].
Conclusion: For binary outcomes, CRTs and IRTs can safely be pooled in MAs because of
an absence of systematic differences between effect estimates. For continuous
outcomes, the results were less clear although accounting for trial sample sizes led to
a non-significant difference. More research is needed for continuous outcomes and,
meanwhile, MAs should be completed with subgroup analyses (CRTs vs IRTs).
VCThe Author(s) 2018. Published by Oxford University Press on behalf of the International Epidemiological Association. 609 This is an Open Access article distributed under the terms of the Creative Commons Attribution Non-Commercial License (http://creativecommons.org/licenses/by-nc/4.0/), which permits non-commercial re-use, distribution, and reproduction in any medium, provided the original work is properly cited. For commercial re-use, please contact firstname.lastname@example.org
doi: 10.1093/ije/dyy229 Advance Access Publication Date: 9 November 2018 Original article
Key words: Cluster randomized trial, individually randomized trial, intervention effect estimate, meta-epidemiologi-cal study, systematic review
Cluster randomized trials (CRTs) are defined as trials in which clusters of participants such as wards, practices, schools or villages are randomized rather than the partici-pants themselves.1These trials are known to be more sus-ceptible to bias than individually randomized trials (IRTs). For instance, recruitment bias may occur when partici-pants are recruited after cluster randomization by a non-blinded recruiter.2–6This situation shares some similarities with the lack of allocation concealment, shown to be asso-ciated with an over-estimation of intervention effects.7In this situation, study groups may be unbalanced in regard to individual baseline characteristics, as the individual is not the unit of randomization.
Some interventions have been assessed with both CRTs and IRTs. In two reviews of hip protectors, large positive effects were seen in CRTs whereas effects in IRTs were more equivocal.8,9Although the intervention assessed may appear simple, Hahn et al.9explained that it actually may differ in two ways between CRTs and IRTs: (i) CRTs may benefit from a ‘herd effect’ with higher compliance; and (ii) IRTs may suffer from inter-group contamination which is often a reason for adopting cluster randomization. Both elements could lead to larger intervention effect estimates in CRTs than in IRTs. Conversely, Gilbody et al.10found similar results between CRTs and IRTs when investigating collaborative care for depression, and Selvaraj and Prasad11found similar proportions of positive results be-tween CRTs and IRTs
These examples remain anecdotal, and to date we lack general findings as to whether intervention effect estimates are, on average, larger in CRTs than in IRTs. It then remains unclear whether these trial types can be pooled in meta-analyses (MAs). The Cochrane Handbook12considers
the unit-of-analysis error for CRTs, but nothing is said regarding a potential systematic difference in intervention effect estimates between these two trial types. Knowing if such a difference exists is, however, crucial for different reasons. First, CRTs and IRTs are often meta-analysed to-gether, but this relies on the assumption that they estimate the same quantity of interest. Second, if there is a system-atic difference between the estimates from the two types of trials, it might suggest that CRTs and IRTs lead to different estimands and therefore the interpretations of the results are different; CRTs keep existing ‘social units’ in which participants can interact. Therefore, CRTs may lead to real-world evidence and estimation of the effectiveness, as opposed to the ‘ideal-world’ estimation of efficacy obtained from IRTs. Third, the presence of systematic dif-ferences would imply that the intervention effect estimates from an IRT could not be used (at least as it is) to inform the sample size of a future CRT and vice versa.
For these reasons, we performed a meta-epidemiological study to assess whether intervention effect estimates are larger in CRTs than in IRTs. With this approach, we aim to understand whether the specificities of the two types of trials lead to systematic differences in intervention effect esti-mates. Indeed, CRTs and IRTs not only differ in the ran-domization procedure but also in ways participants are recruited, the intervention delivered, etc. Hence, we want to quantify the overall impact of these differences on the inter-vention effect estimates. To do so, we compared interven-tion effect estimates for the same interveninterven-tion on the same outcome in studies using cluster randomization and studies using individual randomization. In order to ensure a comparability of the intervention and outcomes, we used tri-als that have been meta-analysed together in systematic reviews, adopting a quantitative approach called a meta-epidemiological study.
• Cluster randomized trials are known to be more pragmatic than individually randomized trials but also more suscepti-ble to bias.
• Cluster randomized and individually randomized trials are often pooled in MAs, but no study has investigated potential systematic differences in intervention effect estimates between these two trial types.
• In MAs of binary outcomes, intervention effect estimates for cluster and individually randomized trials did not differ.
• In MAs of continuous outcomes, intervention effect estimates were moderately more favourable for individually ran-domized trials. However, the difference in intervention effects was moderated by study size and characteristics of out-come and intervention. Therefore, the inconsistent results in subgroup analyses invite further studies.
Meta-epidemiological studies are used to determine which trial characteristics are associated with treatment effect estimates.13 In this study, the characteristic of interest is the design (cluster vs individual randomization). A meta-epidemiological study is generally conducted with a two-step approach using a collection of MAs. First, for each se-lected MA and considering the trial as the unit of analysis, the difference in intervention effect estimates between studies which have the characteristic of interest (i.e. which are cluster randomized, in the present study) and those which are not, is assessed. This is done by fitting one meta-regression for each selected MA. Second, results obtained after this first step are meta-analysed. In this sec-ond step, the units of analysis are the MAs which have ini-tially been selected. Using this approach, our null hypothesis for our primary analysis was an absence of a systematic difference in intervention effect estimates be-tween CRTs and IRTs.
On 10 March 2015, we searched for eligible MAs pub-lished between 1 January 2010 and 31 December 2014 in the Cochrane Database of Systematic Reviews, by using the following keywords: ‘cluster randomized’ OR ‘group randomized’ OR ‘community randomized’ in the full text.
MA and trial selection
Identified systematic reviews were screened to select only those including both CRTs and IRTs. We selected eligible MAs with at least three randomized trials, including at least one CRT and one IRT. Where more than one MA was eligible within the same review, the MA corresponding to the primary outcome, if clearly stated, was selected. Otherwise, the MA with the largest number of trials was selected. We excluded MAs with safety or compliance out-comes and those for which a control group was not clearly identifiable.
Trials were classified as CRT or IRT according to what was reported in the systematic review. Because we were in-terested in comparing trials that used different randomiza-tion units, we discarded quasi-randomized trials and controlled before–after studies. We discarded duplicate tri-als within MAs and kept the duplicate with the largest sample size. We finally discarded duplicate trials between MAs, keeping the duplicate from the most recently pub-lished systematic review. All those steps were performed independently by two of us (C.L., B.G.), with
disagreements resolved by discussion, referring to a third opinion (A.C.) when necessary.
Data extraction and coding
We extracted data related to MAs and trials by using two standardized and pilot-tested spreadsheets. The items we extracted are presented inSupplementary Table 1, avail-able asSupplementary dataat IJE online. Data were col-lected from the systematic reviews, except when otherwise specified. Data extraction was performed inde-pendently by two of us (C.L., B.G.) and any discrepancy was adjudicated, referring to a third opinion (A.C.) when necessary.
If the number of patients per group, number of events, means and standard deviations were not reported in the systematic review, we collected them from the trial reports, or in case of doubt, we contacted the authors of the systematic reviews (which happened for nine MAs).
Accounting for clustering: calculating effective
For CRTs which did not adjust for clustering during analy-sis, we applied the method described in the Cochrane Handbook12; the sample size of the trial was reduced to an effective sample size by dividing the original sample size by the design effect. The design effect is defined as [1 þ (M – 1) q], where M is the average cluster size and q the intra-class correlation coefficient (ICC), the parameter intra- classi-cally used to quantify the clustering effect. We collected the ICC from the trial report and, if not reported, a value of 0.03 was chosen, corresponding to the median ICC value for outcome variables observed in the Campbell et al.14review. A sensitivity analysis doubled this generic value to 0.06 and also considered two extreme situations of no correlation (ICC ¼ 0) and a very strong correlation (ICC ¼ 0.50). If clustering was accounted for, we col-lected the effective sample size reported in the systematic review.
Estimation of intervention effects within each MA
For binary outcomes, intervention effects estimates were expressed as odds ratios. For all outcomes, an odds ratio of less than 1 indicated a beneficial effect of the experimental intervention. For continuous outcomes, intervention effects estimates were expressed as standardized mean differences
using the Hedges and Olkin unbiased estimator of effect size:15 g ¼ 1 3 4 nð 1þ n2 2Þ 1 d;
where n1 and n2 are the size of the control and
experimen-tal group, respectively, and d is the traditional Cohen’s standardized difference: d ¼ ﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃﬃX1 X2 n11 ð Þs2 1þ nð21Þs22 n1n22 q :
X1 and X2 are the observed means in the control and
experimental group, respectively, and s2
1and s22are the two
sample variance estimates. An effect of less than 0 always indicated a beneficial effect of the experimental interven-tion. For each MA, the intervention effect was estimated by using a random effect MA.
We analysed binary and continuous outcomes separately, using the two-step approach proposed by Sterne et al.16 First, for each MA, we performed an inverse-variance weighted random effects meta-regression, thus accounting for between-trial heterogeneity. The only co-variate was the type of trial (cluster or individual randomi-zation), with individual randomization as the reference category. For binary outcomes, we estimated the ratio of odds ratios (ROR), where a ratio of odds ratios less than 1 indicated more favourable intervention effect estimates in cluster trials, meaning that either the intervention was more beneficial or less detrimental in CRTs than in IRTs. For continuous outcomes, we estimated the difference in standardized mean differences (DSMD), where less than 0 indicated more favourable intervention effect estimates in cluster trials. In the second stage, the ratio or difference in intervention effects was combined across MAs using random effects MAs. The heterogeneity between MAs was quantified with the I2, Cochran Q chi-squared
test, and between MAs variance s217using a REML esti-mation.18 Analyses involved use of SAS 9.4 and R 3.2.0 with the package metafor. All the statistical tests were done at a 5% significance level.
Subgroup and adjusted analyses
The type of outcome (objective vs subjective) was a pre-specified subgroup analysis motivated by the fact that Savovic et al.7 observed differences in their meta-meta-epidemiological study according to whether the outcome
was an objective one or not, especially when looking at blinding. The type of intervention (pharmacological vs non-pharmacological) and control intervention (active vs inactive) were post hoc subgroup analyses. For these sub-group analyses, interaction P-values were obtained fitting a random effects meta-regression model with MAs as the unit of analysis and including the variable defining the sub-group. Then, the ROR or DSMD was estimated separately in each subgroup. Planned sensitivity analyses involved adjusting the meta-regression models on each domain of the Risk of Bias tool.19We adjusted the analysis using each item one at a time, considering low vs high or unclear risk. Further post hoc sensitivity analyses were also conducted adjusting on trial sample size. Adjusted analyses were con-ducted excluding MAs with missing data.
Sample size calculation
In order to detect a ratio of odds ratio of 0.85, we required 57 MAs to achieve 80% power using a two-sided 5% sig-nificance level,20assuming a mean number of eight trials per MA21 with an average of three being CRTs, and the following variances: 0.25 for the within-trial variance of the intervention effect estimate; 0.08 for the between-trial within-meta-analysis variance of the intervention effect es-timate; 0.0256 for the between-trial variance of the trial-specific impact of the cluster vs individual randomization; and 0.0016 for the between-meta-analysis variance of the trial-specific impact of cluster randomization. These assumptions were based on the Turner et al.22large epide-miological study of Cochrane MAs. Such a sample size cal-culation supposes a binary outcome. We decided to perform two separate analyses, according to whether the outcome was binary or continuous, and then aimed at identifying at least 57 MAs with a binary outcome.
Characteristics of selected MAs
Considering Cochrane reviews published over a period of five full years, we identified 1301 systematic reviews by the electronic search. We selected 121 MAs (full references in in the Supplementary Material 1a, 1b and 2, available as
Supplementary dataat IJE online), corresponding to 1458 trials (Figure 1): 76 MAs (917 trials) had binary outcomes and 45 (541 trials) had continuous outcomes. MAs con-cerned very different medical and educational fields and interventions (Supplementary Material1a and 1b, available asSupplementary dataat IJE online).
Table 1 shows that pharmacological interventions were investigated in 25 (32.5%) MAs with a binary
outcome but in only six (13.3%) of those with a continu-ous outcome. Less than one-third of the MAs had active controls, both for binary and continuous outcomes. Assessed outcomes were objective in one-third of MAs with a binary outcome and in one-quarter with a continu-ous outcome. The median number of trials (interquartile range: IQR) included was 8 (5 to 15) for MAs with a bi-nary outcome and 10 (5 to 14) for those with a continu-ous outcome. Finally, for MAs with a continucontinu-ous outcome, more than half showed substantial heterogene-ity, as defined in the Cochrane Handbook,12with median I2of 60.4% (IQR 22.8%; 81.7%), whereas for MAs with a binary outcome, the median I2was 26.5% (IQR 0.0%; 53.5%).
Characteristics of selected trials
Among the 917 trials with a binary outcome, 183 (20.0%) were CRTs and 734 (80.0%) were IRTs (Table 1). The median sample size was 570 for CRTs (213 to 1764) and 208 for IRTs (83 to 527). The median number of randomized clusters was 31 (12 to 76) and in half, ran-domized clusters correspond to clinical settings. For 64 of them we used the 0.03 generic value for the ICC to correct the sample size.
Among the 541 trials with a continuous outcome, 131 (24.2%) were CRTs and 410 (75.8%) were IRTs. The median sample size was 139 for CRTs (55 to 291) and 113 for IRTs (56 to 211). Among the 131 CRTs, the me-dian number of randomized clusters was 20 (10 to 40)
Figure 1. Flow diagram of the selection of MAs and randomized trials.
and in less than one-third, randomized clusters corre-sponded to clinical settings (Table 1). For 45 of them we used the 0.03 generic value for the ICC to correct the sample size.
For MAs with a continuous outcome, 30 CRTs (23.4%) were at low risk of bias for blinding of
outcome assessment as compared with 161 IRTs (40.4%), although in most of these, the risk was assessed as unclear (Supplementary Table 2, available as
Supplementary dataat IJE online). For binary outcomes, no difference was observed between CRTs and IRTs in terms of risk of bias.
Table 1. Characteristics of MAs and trials included
MA characteristics MAs with a binary outcome MAs with a continuous outcome
(n ¼ 76) (n ¼ 45)
Intervention, n (%)
Pharmacological 25 (32.9) 6 (13.3)
Non-pharmacological 51 (67.1) 39 (86.7)
Intervention in control group, n (%)
Inactive 52 (68.4) 34 (75.6)
Active 24 (31.6) 11 (24.4)
Outcome objectivity, n (%)
All-cause mortality 14 (18.4)
-Objectively assessed 11 (14.5) 11 (24.4)
Objectively assessed but influenced by clinician or patient 30 (39.5) 5 (11.1)
Subjectively assessed 21 (27.6) 29 (64.4)
Number of trials, median (first and third quartiles) (range)a
Total 8 (5; 15) (3 to 46) 10 (5; 14) (3 to 44)
Cluster randomized trial 2 (1; 3) (1 to 9) 1 (1; 2) (1 to 24) Individually randomized trial 6 (3; 14) (1 to 45) 7 (3; 10) (1 to 38) I2, median (first and third quartiles)a 26.5 (0.0; 53.5) 60.4 (22.8; 81.7)
s2, median (first and third quartiles)a 0.031 (0.000; 0.141) 0.039 (0.005; 0.156)
Binary outcome Continuous outcome
Trial characteristics CRTs IRTs CRTs IRTs
(n ¼ 183) (n ¼ 734) (n ¼ 131) (n ¼ 410) Year of publication, median (first and third quartiles) 2003 (1997; 2008) 2003 (1996; 2007) 2006 (2003; 2009) 2006 (2001; 2009) Sample size,amedian (first and third quartiles) 570 (213; 1764) 208 (83; 527) 139 (55; 291) 113 (56 ; 211)
Mean 6 standard deviation (SD) 7 886 6 43 120 1 589 6 9 059 280 6 424 197 6 354 Cluster type, n (%)
Clinical setting: 94 (51.4) 43 (32.8)
Hospital 12 (6.6) 4 (3.0)
Ward 11 (6.0) 3 (2.3)
Health centre 13 (7.1) 1 (0.8)
Residential care home 10 (5.5) 5 (3.8)
Practice or health professional 43 (23.5) 24 (18.3)
Other 5 (2.7) 6 (4.6)
Non-clinical setting: 85 (46.4) 87 (66.4)
School or classroom 20 (10.9) 63 (48.1)
Family/household 12 (6.6) 4 (3.0)
Village or geographical area 37 (20.2) 5 (3.8)
Other 16 (8.7) 15 (11.5)
Unclear 4 (2.2) 1 (0.8)
Number of clusters, median (first and third quartiles), (range) 31 (12; 76), 20 (10; 40), (2 to 68 146) (4 to 531)
aFor CRTs, sample size was corrected for clustering.
CRT: cluster randomised trial, IRT: individually randomized trial.
Figure 2. Differences in intervention effect estimates between cluster and individually randomized trials with a binary outcome.
Differences in intervention effect estimates
between CRTs and IRTs
For MAs with a binary outcome, intervention effect esti-mates did not differ between CRTs and IRTs. The com-bined ROR was estimated at 1.00 (95% CI 0.93 to 1.08) (Figure 2andTable 2). Heterogeneity was low across MAs (I2 ¼ 21.2%; P ¼ 0.238; between-meta-analyses variance s2¼0.018). Subgroup and adjusted analyses led to consis-tent results with a combined ROR very close to 1.00, whatever the analysis (Table 2andSupplementary Figures 2–4, available as Supplementary dataat IJE online). The results were also robust across all the performed sensitivity analyses (see Supplementary Tables 3–5, available as
Supplementary dataat IJE online).
For MAs with a continuous outcome, intervention ef-fect estimates were more favourable for IRTs, with a com-bined DSMD of 0.13 (95% CI 0.06 to 0.19) (Figure 3). Although statistically significant, this different is small according to Cohen’s classification of effect sizes23 Heterogeneity was low across MAs (I2 ¼ 21.7%; P ¼ 0.221; between-meta-analyses variance s2¼0.009).
Subgroup analyses led to inconsistent results among subgroups. The combined DSMD was significant for pharmacological interventions but was lower and non-significant for pharmacological interventions: 0.15 (0.08 to 0.21) vs -0.03 (-0.12 to 0.07) (interaction P-value ¼ 0.016). Similarly, the effect of cluster randomiza-tion on intervenrandomiza-tion effect estimates was larger for subjec-tive than for objecsubjec-tive outcomes, although the interaction
Table 2. Difference in intervention effect estimates between cluster and individually randomized trials for binary and continuous outcomes
Binary outcome n* Estimate 95% CI P-value I2(95% CI) s2(95% CI) P-value
Global 76 1.00 (0.93 to 1.08) 0.238 21.2 (0.0 to 41.4) 0.018 (0.000 to 0.047) Subgroup analyses Pharmacological 25 1.02 (0.94 to 1.10) 0.405 0.0 (0.0 to 64.9) 0.000 (0.000 to 0.103) 0.360 Non-pharmacological 51 0.98 (0.89 to 1.08) 0.218 20.9 (0.0 to 43.9) 0.023 (0.000 to 0.067) Subjective 51 0.99 (0.89 to 1.10) 0.090 26.9 (0.0 to 48.9) 0.035 (0.000 to 0.902) 0.496 Objective 25 1.00 (0.93 to 1.08) 0.738 0.0 (0.0 to 49.3) 0.000 (0.000 to 0.047) Active 24 1.02 (0.89 to 1.15) 0.657 6.0 (0.0 to 43.0) 0.006 (0.000 to 0.072) 0.929 Inactive 52 1.01 (0.91 to 1.11) 0.114 29.0 (0.0 to 57.6) 0.025 (0.000 to 0.083) Adjusted on risk of bias of:
Generation of random sequence 60 1.03 (0.94 to 1.12) 0.005 37.8 (5.6 to 63.6) 0.034 (0.003 to 0.097) Allocation concealment 60 1.01 (0.92 to 1.11) 0.012 37.1 (2.6 to 59.2) 0.034 (0.002 to 0.084) Blinding for participants 31 1.01 (0.87 to 1.18) 0.001 53.0 (14.2 to 80.1) 0.062 (0.009 to 0.220) Blinding for the outcome assessor 44 0.99 (0.89 to 1.11) 0.014 40.8 (2.6 to 63.0) 0.040 (0.002 to 0.099) Adjusted on trial sample size 69 0.98 (0.89 to 1.07) 0.056 27.1 (0.0 to 58.3) 0.028 (0.000 to 0.105)
Continuous outcome n* Estimate 95% CI P-value I2(95% CI) s2(95% CI) P-value
Global 45 0.13 (0.06 to 0.19) 0.221 21.7 (0.0 to 47.4) 0.009 (0.000 to 0.029) Subgroup analyses Pharmacological 6 0.03 (-0.12 to 0.07) 0.435 0.0 (0.0 to 90.6) 0.000 (0.000 to 0.436) 0.016 Non pharmacological 39 0.15 (0.08 to 0.21) 0.515 7.5 (0.0 to 43.2) 0.003 (0.000 to 0.027) Subjective 34 0.15 (0.08 to 0.22) 0.398 11.1 (0.0 to 52.5) 0.005 (0.000 to 0.040) 0.118 Objective 11 0.05 (-0.08 to 0.17) 0.420 20.5 (0.0 to 74.2) 0.008 (0.000 to 0.091) Active 11 0.25 (0.15 to 0.36) 0.877 0.0 (0.0 to 57.2) 0.000 (0.000 to 0.049) 0.006 Inactive 34 0.08 (0.01 to 0.15) 0.352 15.4 (0.0 to 54.5) 0.006 (0.000 to 0.037) Adjusted on risk of bias of:
Generation of random sequence 32 0.12 (0.05 to 0.19) 0.583 8.8 (0.0 to 48.0) 0.003 (0.000 to 0.031) Allocation concealment 36 0.11 (0.03 to 0.19) 0.116 29.3 (0.0 to 60.9) 0.013 (0.000 to 0.050) Blinding for participants 16 0.11 (0.00 to 0.22) 0.065 38.3 (0.0 to 78.9) 0.016 (0.000 to 0.094) Blinding for the outcome assessor 23 0.22 (0.03 to 0.41) <0.0001 84.6 (66.7 to 93.0) 0.134 (0.049 to 0.328) Adjusted on trial sample size 38 0.06 (-0.02 to 0.13) 0.060 24.1 (0.0 to 75.1) 0.011 (0.000 to 0.102)
n*, number of MAs included in the analysis.
was not significant: 0.15 (0.08 to 0.22) vs 0.05 (-0.08 to 0.17) (interaction P-value ¼ 0.118). Finally, the DSMD was significantly lower for inactive compared with active control interventions: 0.08 (0.01 to 0.15) vs 0.25 (0.15 to 0.36) (interaction P-value ¼ 0.006). Adjusting for the effective trial sample size led to a smaller difference of 0.06 (-0.02 to 0.13), which was not significant. Adjusting for risk of bias items did not affect the results, except for blinding of outcome assessors, with a higher DSMD, estimated to be 0.22 (0.03 to 0.41). The choice of the ICC value to adjust for clustering when the trials values were not known does not impact the results (results are presented in Supplementary Table 3, available as
Supplementary dataat IJE online).
In this meta-epidemiological study, we selected 121 MAs: 76 (917 trials) with a binary and 45 (541 trials) with continuous
outcomes. For binary outcomes, the ratio of odds ratios was 1.00 (95% CI 0.93 to 1.08), indicating that intervention ef-fect estimates did not systematically differ between CRTs and IRTs. Consistent results were observed in all subgroup and adjusted analyses. For continuous outcomes, interven-tion effect estimates were more favourable with individual randomization, although the difference was moderate (dif-ference in standardized mean dif(dif-ferences of 0.13, 95% CI 0.06 to 0.19). This difference was much smaller and not sig-nificant for the trial subgroup of pharmacological interven-tions or when adjusting on sample size.
Strengths and limitations of the study
We selected a large sample of MAs covering a wide range of medical and educational areas, which provides good gener-alizability of our results. We nevertheless restricted our study to Cochrane MAs: to identify potentially eligible MAs, we had to access the full text of the systematic reviews
Figure 3. Differences in intervention effect estimates between cluster and individually randomized trials with a continuous outcome.
because the abstracts of reports rarely specify the inclusion of both CRTs and IRTs. Restricting our study to Cochrane reviews may limit the generalizability of our results. This study was conducted using trial-level summaries of the in-tervention effect. Therefore, no information was available regarding patients’ non-adherence or loss to follow-up, which might have had an impact on the trials’ results, if these issues were to affect CRTs differently from IRTs. However, our aim was to assess whether there exists sys-tematic differences between CRTs and IRTs. Further studies using individual patient data would be needed to investigate the specific effect of each component that differs between CRTs and IRTs. Such studies would probably need to re-strict the focus to a specific medical area, which differs from the philosophy of meta-epidemiological studies.
We discarded studies not randomized, such as quasi-randomized trials or before–after studies, so as to obtain well-defined groups for comparison. We handled cluster-ing, thus making sure that our results were not distorted by over-weighted CRTs. Finally we explored both binary and continuous outcomes in the same study (although indepen-dently) which, except for the Alexander et al.24 or Smaı¨l-Faugeron et al.25studies, is uncommon.
Relation to previous work
To our knowledge and in view of the Dechartres et al.26 re-cently published systematic review of meta-epidemiological studies, our study is the first meta-epidemiological study to compare intervention effect estimates between CRTs and IRTs. However, our results are consistent with Selvaraj and Prasad’s,11who showed that the proportions of statistically significant findings were similar in CRTs and IRTs.
CRTs and IRTs differ in several ways. CRTs may face recruit-ment bias, but they may benefit from a ‘herd effect’; IRTs may suffer from group contamination. All these elements may lead to larger intervention effect estimates in CRTs than in IRTs. Besides, most of the interventions assessed in CRTs do not al-low for any form of blinding, which invites both performance and detection bias.6This feature has been shown to be
associ-ated with an over-estimation of intervention effects.7 Conversely, CRTs are considered more pragmatic,27and al-low the estimation of the effectiveness, rather than the effi-cacy, as in many IRTs. Effectiveness is usually smaller than efficacy, mainly because of non-compliance. Pragmatic trials also nearly always involve several centres and they are usually larger. These characteristics are important because the inter-vention effect estimates have been shown to be lower in multi-centre than single-multi-centre trials,28,29and in larger trial sample
sizes.30 Therefore, antagonist mechanisms may occur and might counterbalance each other. In the end, although CRTs and IRTs may look as if they are similar but just conducted as CRTs or IRTs, very different mechanisms—sometimes antag-onist—may apply and contribute to systematic differences in intervention effect estimates between CRTs and IRTs.
Discrepancy between binary and continuous
The finding that there is no difference between CRTs and IRTs for binary outcomes suggests that the different mecha-nisms are not very strong, or non-existent or that they compen-sate for each other, and this result held in all considered subgroups. For continuous outcomes, the observed 0.13 differ-ence in standardized mean differdiffer-ences invites the two following comments. First, although significant, the observed difference can be considered moderate in view of previously reported dif-ferences in standardized mean difdif-ferences.26Second, one could have expected a difference in the opposite sense in view of the underlying mechanisms (i.e. larger intervention effect estimates in CRTs than in IRTs). A potential explanation is that there are probably many single-centre IRTs, with low median size (113 participants), whereas CRTs are intrinsically multicentre studies, most randomizing practices, schools or classrooms.
The discrepancy we observed between MAs with binary and continuous outcomes is not new, and others have urged caution when extrapolating results of meta-epidemiological studies of binary outcomes to situations of continuous out-comes.24,26We found several differences between trials and MAs according to whether the outcome was continuous or binary: (i) the sample size was smaller in trials with continu-ous outcomes; (ii) heterogeneity was higher (median I2 of
60.4 compared with 26.5); (iii) blinding was less frequent; (iv) outcomes were more frequently subjective (64.4% of MAs with a continuous outcome vs 27.6% with a binary outcome when focusing on only ‘subjective outcome’; and (v) the settings differed, with cluster trials with a continuous outcome being more likely to have non-clinical settings. All these differences may explain the discrepancy we observed.
Finally, from a statistical point of view, we cannot ex-clude some form of meta-confounding. We indeed adjusted analyses, but doing so led to discarding some MAs (nota-bly those with only three trials), and we adjusted on only one covariate at a time.
Conclusions and implications
For binary outcomes, CRTs and IRTs produced the same intervention effect estimates, but intervention effect esti-mates were marginally more favourable (i.e. either more beneficial or less detrimental) for IRTs with continuous outcomes. However, this result was not observed for trials
assessing a pharmacological intervention or with an objec-tive outcome. More work is needed, in particular to under-stand how the type of intervention, outcome, setting or trial sample size affects the results.
Supplementary dataare available at IJE online.
This work was supported by a grant from the French Ministry of Health (PREPS 13-0015).
Conflict of interest: None declared.
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