We used the NHS observational data to emulate the design and analysis of the WHI randomized trial. The ITT (intention-to-treat) hazard ratios of CHD for therapy initiation were 1.42 (CI = 0.92–2.20) in the NHS versus 1.68 (95% CI = 1.15–2.45) in the WHI9
during the first two years, and 1.00 (0.78–1.28) in the NHS versus approximately 1.24 (0.97–1.60) in the WHI8
during the first 8 years. However, much of the apparent WHI-NHS difference disappeared after stratification by time since menopause at hormone therapy initiation. The ITT hazard ratios were 0.84 (0.61–1.14) in the NHS versus 0.88 (0.54–1.43) in the WHI8, 18
for women within 10 years after menopause, and approximately 1.12 (0.84–1.48) in the NHS versus 1.23 (0.85–1.77) in the WHI8, 18
for women between 10 and 20 years after menopause.
These findings provide additional support to the hypothesis that hormone therapy may increase the long-term CHD risk only in women who were 10 or more years after menopause at initiation,17, 19
and to the rationale for an ongoing randomized clinical trial to determine the effect of estrogen plus progestin on coronary calcification in younger women.20
When the analyses were limited to women with no past history of hormone use, the ITT analysis result was 0.79 (0.60–1.03) for the entire follow-up and 0.66 (0.44–0.98) for women who initiated hormone use within ten years of menopause.
We computed average ITT hazard ratios in the NHS for comparison with the main result of the WHI. Our ITT estimates suggest that any remaining differences between NHS and WHI estimates are not explained by unmeasured joint risk factors for CHD and therapy discontinuation. However, the average ITT hazard ratio is not the ideal effect measure because the survival curves crossed during the follow-up in both the WHI trial and the NHS “trials,” and also because ITT estimates like the ones shown here are generally attenuated towards the null due to misclassification of actual treatment. We addressed the problem by estimating survival curves to first CHD event, and by estimating these curves under full adherence (via inverse probability weighting). Therefore the adherence-adjusted survival curves of provide the most appropriate summary of our results. It will be of interest to compare these results with adherence-adjusted curves (via inverse probability weighting) from the WHI when they become available. The curves suggest that continuous hormone therapy causes a net reduction in CHD among women starting therapy within 10 years of menopause, and a net increase among those starting later. However, either of these effects (as well as the qualitative interaction between therapy and time since menopause) could be due to sampling variability.
Previously published NHS estimates17
compared the hazards of current versus never users over the two-year period after the updating of treatment status at the return of each questionnaire, and could thus be viewed as a form of adherence adjustment. In we described the steps from our two-year ITT estimate to the previously published adherence-adjusted estimate. Below we discuss the two key steps: the change of start of follow-up (time of therapy initiation vs. time of questionnaire return), and the change of the exposed group (selected initiators vs. current users).
The two-year hazard ratio estimate changed from 1.42 () to 0.98 (, column i) during the first two years, and from 0.96 () to 0.84 (, column i) for the entire follow-up when the definition of start of follow-up was changed from the estimated time of therapy initiation to the time of return of the next questionnaire. (The latter definition is commonly used in observational studies that collect treatment information at regular intervals.) This latter definition excludes women who initiated treatment and then suffered a nonfatal MI during the interval between treatment initiation and treatment ascertainment (up to two years in the NHS). If hormone therapy increases the short-term risk of CHD, this exclusion will result in an underestimate of the early increase in risk and may result in selection bias,16
which may explain part of the change from 1.42 to 0.98. The impact of this exclusion bias, however, will be diluted over the entire follow-up, as previously suggested in a sensitivity analysis.17
This may explain the smaller change from 0.96 to 0.84. This exclusion bias may be quantified through simulations,21
reduced by stratification of the analysis on duration of therapy at baseline,21
and eliminated by making the start of follow-up coincident with the time of treatment initiation, as discussed by Robins22, 23
The approach we present here and elsewhere10, 25
generalizes Ray’s “new-users design” to the case of time-varying treatments.
The point estimate further changed from 0.98 (, column i) to 0.77 (column ii) when the definition of exposure changed from selected initiators to current users. These are estimates for different contrasts. The estimate in column (i) is based on the exposed person-time during the two-year period immediately after the return of the questionnaire in which therapy initiation was reported, and thus can be viewed as a flawed attempt to estimate the early effect of therapy initiation (see previous paragraph). The estimate in column (ii), however, is based on the exposed person-time pooled over all two-year periods after the return of any questionnaire, and thus can be interpreted as an attempt to estimate the effect of therapy use during any two-year period (that excludes the interval between therapy initiation and return of the next questionnaire, as discussed in the previous paragraph). More specifically, the approach in column (ii) can be understood as an attempt to estimate adherence-adjusted effects by entering the current value of exposure and the joint predictors of adherence and CHD as time-varying covariates in the model for CHD risk. Unlike inverse probability weighting, this approach to adherence adjustment requires that the time-dependent covariates not be strongly affected by prior treatment. This may be a reasonable assumption in the NHS. Thus the estimates in column (ii) may be more usefully compared with a weighted average of our interval-specific adherence adjusted estimates of 1.61 (0–2 years), 0 .65 (2–5 years), 0.47 (5–8 years), and 0.85 (>8 years) than to the estimate in column (i).
In summary, our findings suggest that the discrepancies between the WHI and NHS ITT estimates could be largely explained by differences in the distribution of time since menopause and length of follow-up. Residual confounding for the effect of therapy initiation in the NHS seems to play little role.