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Statistical Methods in Medical Research 4ed

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1993; Prosk<strong>in</strong>, 1993). There is thus scope for overviews of related meta-analyses<br />

(M 2 -analyses?), with the loom<strong>in</strong>g possibility of M 3 -analyses, and so on!<br />

Analysis<br />

A useful start<strong>in</strong>g-po<strong>in</strong>t is the method of weight<strong>in</strong>g described <strong>in</strong> §8.2. The difference<br />

<strong>in</strong> efficacy between two treatments will be measured by a suitable parameter,<br />

typically chosen as the difference <strong>in</strong> mean responses or (for b<strong>in</strong>ary<br />

responses) the difference <strong>in</strong> proportions or the log odds ratio. Suppose there<br />

are k trials, and that for the ith trial the true value of the difference parameter is<br />

ui, estimated by the statistic yi, with estimated sampl<strong>in</strong>g variance vi. Def<strong>in</strong>e the<br />

weight wi ˆ 1=vi. Then, if all the ui are equal to some common value u, a suitable<br />

estimate of u is the weighted mean<br />

P<br />

wiyi<br />

y ˆ P , …18:6†<br />

wi<br />

as <strong>in</strong> (8.13), and, approximately, var…y† ˆ1= P wi. The assumption that the<br />

ui are constant may be tested, as <strong>in</strong> (8.14) and (8.15), by the heterogeneity<br />

statistic<br />

G ˆ P wi…yi y† 2 ˆ P wiy 2 P<br />

i … wiyi†<br />

2 = P wi, …18:7†<br />

distributed approximately as x 2 …k 1† .<br />

18.10 Meta-analysis 643<br />

For trials with a b<strong>in</strong>ary outcome, these formulae may be applied to provide<br />

p<br />

comb<strong>in</strong>ed estimates of log relative risk (us<strong>in</strong>g (4.24) for vi)<br />

or log odds ratio<br />

(us<strong>in</strong>g (4.26). In the latter case, an alternative approach is to use the method of<br />

Yusuf et al. (1985) given <strong>in</strong> (4.33). If Oi and Ei are the observed and expected<br />

numbers of critical events for one of the treatments, the log odds ratio is<br />

estimated by yi ˆ…Oi Ei†=Vi, with var…yi† ˆ1=Vi, where Vi is calculated, as<br />

<strong>in</strong> (4.32), from the 2 2 table for the trial. Then the weighted mean (18.6)<br />

becomes<br />

P<br />

…Oi Ei†<br />

y ˆ P ,<br />

Vi<br />

with var…y† ˆ1= P Vi. The heterogeneity statistic (18.7), distributed approximately<br />

as x2 …k 1† , becomes<br />

G 0 ˆ P …Oi Ei† 2<br />

‰<br />

Vi<br />

P …Oi Ei†Š 2<br />

P :<br />

Vi<br />

As noted <strong>in</strong> §4.5, this method is biased when the treatment effect is large (i.e.<br />

when the odds ratio departs greatly from unity). The report by the Early Breast<br />

Cancer Trialists' Collaborative Group (1990) suggests that the bias is not serious

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