To the Editor: Problems with interpreting odds ratios reported in meta-analyses of smoking-cessation interventions have recently been highlighted.1 Ford and Dobson2 have erred in a different way when applying the findings of the Cochrane review on smoking cessation interventions in pregnancy3 to calculate the public health benefits of delivering such interventions to all pregnant women in Australia.

When all methodologically acceptable randomised controlled trials were considered, the Cochrane review did find the prevalence of smoking at end-of-pregnancy was 6% lower in intervention than control groups.3 However, this does not equate to a 6% reduction in the population prevalence of smoking among pregnant women, as Ford and Dobson assume. A mean between-group difference reported in a meta-analysis is not equivalent to a difference of exactly the same magnitude in a population prevalence of a risk factor unless 100% of the population exhibit that risk factor. Clearly, as Ford and Dobson have reported, this is not the case with smoking in pregnancy, where they correctly note that about 20% of pregnant women report current smoking at their first antenatal visit.2 Therefore, smoking-cessation interventions would not reduce the prevalence of smoking by 6% from 20% to 14%. The expected reduction can be calculated as follows: expected reduction in prevalence of smoking in pregnant women = current smoking prevalence in pregnant women (20%) × between-group difference in smoking prevalence (0.06) = 1.2%.

This calculation rests on two assumptions: namely, that all pregnant women in Australia currently receive usual smoking-cessation care equivalent to that of control group conditions in the Cochrane review3 and that, in the short term, antenatal care can be transformed to the point where all future pregnant women receive smoking-cessation care equivalent to that received by those in intervention groups in the Cochrane review.

Therefore, it is obvious that the expected smoking prevalence of 18.8% (20% minus 1.2%) is considerably higher than the 14% calculated by Ford and Dobson.2 Unfortunately, this means the rates of reduced infant deaths, hospital separations and costs to the healthcare system estimated by Ford and Dobson have also been overstated.

In summary, the gains to be expected by clinical interventions with pregnant smokers are modest. Furthermore, past evaluations of media campaigns directed specifically at pregnant women have not shown significant positive effects.4 This reinforces the importance of tobacco-control strategies which target the whole population in addition to those which target pregnant women.5

In reply: We thank Walsh and Lumley for correcting the error in our letter. The 6% reduction in smoking during pregnancy referred to an absolute difference in prevalence of continued smoking in late pregnancy among women who smoked early in pregnancy, from 91% in the control groups to 85% in the treatment groups.1 We incorrectly hypothesised a reduction from 20% to 14% in prevalence of any smoking during pregnancy.

In fact, there was a decline in smoking during pregnancy, from 22% in 1994 to 17% in 2001 in New South Wales.2 These figures illustrate well the final point that Walsh and Lumley make: whole-of-population approaches to smoking reduction can yield much greater benefits (a 5% reduction in 7 years in NSW) than high-risk approaches (from our data, the 1.2% calculated by Walsh and Lumley).

Our estimates of the adverse effects of smoking in pregnancy are, at present, correct. Although we unfortunately overstated the possible reductions resulting from interventions targeted only at pregnant women, such reductions are plausible for whole-of-population approaches.3