Timing of Menarche and First Birth in Relation to Risk of Breast Cancer in A-Bomb Survivors

Background

Early age at menarche and late age at first birth are two well-established breast cancer risk factors (1, 2). Between these two milestones, breast tissue is largely undifferentiated and is thus particularly susceptible to carcinogenic insults (3). It is well established that pregnancy, particularly pregnancy at a young age, reduces the risk of breast cancer (4), and this is primarily because of the differentiation of breast tissue that occurs during pregnancy (5). Circulating hormones released during the menstrual cycle can have a greater effect on the promotion of breast abnormalities on the nulliparous, relatively undifferentiated breast compared with the parous breast (4). Thus, the longer the interval between menarche and first birth, an interval Russo et al. (5) have described as a “window of high susceptibility,” the more a woman's breast tissue is particularly susceptible to carcinogens. A few studies have shown a positive association between the timing of menarche and first birth and breast cancer risk (6, 7). Specifically, Li et al. (6) found a 1.5-fold increase in the risk of breast cancer for women who had an interval of ≥16 years between the ages of menarche and first birth in comparison with those who had ≤5 years between these ages, and this risk did not vary substantially by menopausal status.

Radiation exposure is another well-established breast cancer risk factor (8, 9). The purpose of this study is to assess how radiation exposure during the interval between menarche and first birth affects breast cancer risk, compared with those exposed to radiation either before experiencing menarche or after their first birth, among female survivors of the atomic bombings of Hiroshima and Nagasaki in 1945. We hypothesize that the radiation-related risk of breast cancer is higher in women exposed to radiation during the period between menarche and first birth, compared with women of the same age exposed either before menarche or after their first birth.

Evidence exists from previous studies of atomic bomb survivors that age at exposure to radiation is strongly associated with the subsequent risk of breast cancer, with women exposed before age 20 years having a significantly higher radiation-related excess risk before age 35 years than those exposed later (10). In other irradiated populations, evidence suggests that radiation exposure during a first pregnancy was more strongly associated with an elevation in breast cancer risk than radiation exposure during a later pregnancy (11). However, a recent study of breast cancer risk in a cohort of United States women exposed to diagnostic X-ray radiation showed no significant difference in risk associated with the radiation dose-response between women exposed before breast budding, between budding and menarche, between menarche and first live birth, and after first live birth (12). Previous studies using data from the Life Span Study (LSS) and Adult Health Study (AHS) cohorts of atomic bomb survivors have shown that younger age at exposure is associated with an increased risk of breast cancer, particularly exposure before age 20 years (10). This finding, if true, might be due to the inclusion of many women who were exposed between menarche and first birth in this younger age group. However, the most recent incidence study found that the strength and statistical significance of the estimated effect of age at the time of the bombings (ATB) on radiogenic risk estimates described by Land et al. (10) is highly dependent on the modeling of the background rates (13).

Results

During the 904,078 migration-adjusted person-years of follow-up accumulated between 1958 and 2002, a total of 641 breast cancer cases meeting the criteria for inclusion were diagnosed in 30,113 cohort members who could be classified by reproductive status. At the time of the bombing 9,128 (30%) women were premenarche; 7,149 (24%) were between menarche and a first birth; and 13,836 (46%) were post–first birth. Table 1 describes these women and the 30,865 in the LSS and AHS cohort who were not excluded by the criteria above but for whom reproductive status could not be determined. A higher percentage of women between menarche and first birth ATB received high doses of radiation, with 4.6% of women in this reproductive status group receiving a dose of ≥1 Gy compared with only 2.9% of women who were premenarche and 2.6% of women who were post–first birth. Women who were post–first birth ATB tended to have older ages at menarche and younger ages at first birth compared with women in either the premenarche or between menarche and first birth reproductive status groups. Specifically, nearly 70% of women who were post–first birth ATB were 15 years or older at menarche compared with 46.4% of women between menarche and first birth ATB, and 44.3% of women premenarche ATB. Over 80% of women who were post–first birth ATB were ages <25 years at their first birth, whereas fewer than 60% of women in the premenarche or between menarche and first birth groups were ages <25 years at their first birth. Women between menarche and a first birth were much more likely to be nulliparous (5.2%), compared with women who were premenarche ATB (0.5%).

Compared with women who could be classified into one of the three reproductive status groups, women who were not included in the analyses of breast cancer risk because of missing data tended to be older ATB; 20.4% were ≥50 years of age ATB. Women with missing data were also more likely to be NIC and have a radiation dose estimate <0.05 Gy, compared with women in each reproductive status category.

To investigate the primary hypothesis of dose effect modification by reproductive status, the baseline risk of breast cancer was first modeled to depend on an individual's city of residence, age ATB, and attained age, as previously described. Overall, the relative risk (RR) increased with increasing radiation dose, and the ERR associated with a 1-Gy increase in radiation exposure was 1.55 (95% CI, 0.87-2.42) for a woman who was age 30 years ATB and reached an attained age of 70 years (Table 2). In this linear ERR model, the ERR/Gy is allowed to vary as a function of age ATB and attained age. However, neither the observed 3% (95% CI, −31 to 36%) decrease in risk with each decade increase in age at exposure (age ATB) nor the decrease in proportion to the 0.78th power of attained age (95% CI for exponent of attained age: −2.4 to 0.9) significantly affect the ERR/Gy. The estimated EAR was 12.64 cases per 10,000 person-year-Gy, again for age 30 years ATB and attained age 70 years. A significant decrease in EAR with each decade increase in age ATB of 43% was observed (95% CI, −57 to −26%). On the additive scale, the risk of breast cancer increased in proportion to the 2.8th power of attained age (95% CI, 1.7-3.9). When the radiation-associated excess risk of breast cancer was allowed to vary by reproductive status as in Table 3, marginally significant effect modification was observed based on likelihood ratio tests for the significance of the categorical excess risk per Gy terms (ERR: χ² = 6.04, df = 2, P = 0.049; EAR: χ² = 7.09, df = 2, P = 0.029). Using age 30 years ATB and attained age 70 years as a common point of reference for purposes of comparison, women between menarche and first birth have an ERR (2.36/Gy; 95% CI, 1.12-4.13) nearly twice as large as the ERR of women who were premenarche (1.31/Gy; 95% CI, 0.36-3.58) and women who were post–first birth (1.10/Gy; 95% CI, 0.41-2.07; Table 3). Similar dose effect modification was observed in the EAR model with 19.01 (95% CI, 10.02-31.0) excess breast cancer cases per 10,000 per person-year per Gy in women between menarche and a first birth, compared with an excess of only 9.56 (95% CI, 2.91-24.60) among premenarchal women and 2.82 (95% CI, 1.71-3.92) among post–first birth women.

To investigate this dose effect modification further, we sought to account for possible heterogeneity in the baseline risk of breast cancer between reproductive status groups. Allowing the baseline risk of breast cancer to vary with reproductive status, in addition to city, age ATB, and attained age, significantly improved the fit of the models (ERR: χ² = 16.8, df = 2, P < 0.001; EAR: χ² = 15.1, df = 2, P < 0.001). The addition of reproductive status to the baseline model, while improving overall fit, did not substantially change the overall radiation-associated risk of the simpler model that did not include reproductive status in the baseline model. Specifically, the overall ERR per Gy when reproductive status was included in the baseline model was 1.44 (95% CI, 0.79-2.29; Table 4), compared with 1.55 (95% CI, 0.87-2.42) when it was not. More strikingly, the variation in risk across the reproductive status categories was no longer present when reproductive status was incorporated into the baseline model. Neither the ERR nor the EAR model showed evidence of statistically significant dose effect modification by reproductive status (ERR: χ² = 0.25, df = 2, P = 0.88; EAR χ² = 2.7, df = 2, P = 0.26; Table 5).

All analyses were repeated excluding women who were older than 50 years ATB in an effort order to exclude women who were likely to be postmenopausal at the time of exposure. The results of the analyses excluding these older women did not differ significantly from the analyses presented above (data not shown). All analyses were also repeated excluding nulliparous women to examine whether the results were due to an effect of timing of exposure or an effect of parity. The results of these analyses excluding nulliparous women did not differ significantly from the analyses presented above (data not shown). In addition, we modeled the length of the interval between menarche and a first pregnancy (in years) in the background rate of breast cancer (data not shown). However, doing so did not improve the model fit beyond controlling for reproductive status. To investigate the robustness of the results to the background models, we also used categorical representations of age ATB and attained age in the background model in place of the continuous forms of age ATB and attained age, which did not markedly affect the results (data not shown). Finally, a model was fit in which the modifying effects of age at exposure and attained age were allowed to differ for each reproductive status group, but again, this model did not substantially improve the fit of the model in which there was a single modifying effect for age at exposure and attained age.

Discussion

This is the first article to use the atomic bomb survivor cohort to examine whether the interval between menarche and a first pregnancy is a particularly sensitive period in a woman's life in relation to the risk of radiogenic breast cancer. The well-defined LSS and AHS cohort provides an important source of data to address this issue.

After accounting for significant heterogeneity in the baseline risk of breast cancer among women in different reproductive status groups, we did not observe a significant difference in the risk of breast cancer among women exposed to radiation between menarche and a first pregnancy compared with women exposed either before menarche or after a first birth. Although we did not observe the anticipated effect modification by reproductive status, we did observe a significant association between being in this sensitive window ATB and the baseline risk of breast cancer. In models in which the baseline risk of breast cancer was taken to depend on reproductive status, compared with women who were premenarchal ATB, women between menarche and a first birth ATB had a 37% (95% CI, 0-88%) higher background risk of breast cancer, and women who were after first birth had a 10% reduced risk of breast cancer compared with the premenarche ATB group (data not shown). This association between reproductive status and the baseline breast cancer was independent of the radiation effect.

It is likely that women with longer intervals between menarche and first birth had a higher probability of being between menarche and first birth at the time of exposure. This can be thought of as a type of wait time bias, whereby women with a longer interval have a higher probability of being sampled at any given time than women with a shorter interval. In addition to wait time bias, the length of the interval between menarche and first birth may be longest for women who were between menarche and first birth ATB as a consequence of the war. Given the clear biological evidence that this window is relevant because of the susceptibility of undifferentiated breast tissue (5, 6), women with a longer interval between menarche and first birth would be expected to have a higher baseline risk of breast cancer. This would contribute to the observed significance of reproductive status on the baseline risk of breast cancer in our results.

Further, it is likely that radiation induces breast carcinogenesis to some extent regardless of timing of exposure given the increased radiation-associated risk at all ages of exposure, noted in our results and those of previous studies (10, 11, 13). Although the timing of menarche and first birth may be associated to breast cancer risk through the effect of endogenous hormones on undifferentiated breast tissue, the pathway by which radiation causes damage may be independent and less affected by these reproductive milestones. This may explain why statistically significant dose effect modification by reproductive status was observed only when reproductive status was not included in the baseline risk, and that upon accounting for heterogeneity across reproductive status groups by adding it to the baseline risk model, the dose effect modification was no longer significant.

Our estimates of the radiation associated breast cancer risk are generally consistent with the estimates given in the most recent incidence report on solid cancers in the atomic bomb survivors (13). Our hypothesis was that women exposed to radiation between menarche and first birth will have a higher radiation-related breast cancer risk than women exposed at the same age but outside of this interval. Thus, it refers to an effect that can operate in addition to the age ATB effect. Therefore, our findings are not necessarily inconsistent with the observed early-onset effect described by Land et al. (10). Comparing our overall ERR and EAR results to that report, the combined effects of radiation, age ATB, and attained age are similar, as evidenced by the overlapping CIs. However, our estimate of 12.64 cases per 10,000 person-years per Gy (95% CI, 8.18-17.7) is somewhat larger than the estimate of 9.2 cases per 10,000 person-years per Gy (95% CI, 6.8-12.0). This suggests that the effect of radiation on an additive scale may differ in this subset of women from both the LSS and AHS with complete data on reproductive status, compared with the entire cohort used in the incidence analysis. The higher ERR and EAR estimates observed in this study, compared with studies using the entire female LSS cohort, may be due in part to the younger overall age of this subcohort.

This study has several limitations. Unlike articles on breast cancer incidence in atomic bomb survivors that include the entire female LSS cohort, this study was limited to women with available data on reproductive status ATB. Of the 70,146 women in the LSS cohort, only 44,024 completed at least one LSS questionnaire or AHS clinical exam. Between missing data and our exclusion criteria, the number of women available for analysis was reduced to 30,113. Insofar as this subset of women with complete data differed from the cohort as a whole, the results presented here may not be generalizable to all female atomic bomb survivors. We attempted to investigate the differences between women included in the analysis and those who were excluded due to missing data on reproductive status. The main difference between those included and those with missing reproductive status information was that the excluded women tended to be older ATB. We did not attempt to assign women with missing data to a reproductive status category based solely on their age ATB because we felt that doing so might bias the results. However, the data that we do have on the 30,865 women excluded because of missing data on reproductive status, and not due to other prespecified exclusions, described in Table 1, suggest that there are no systematic differences between excluded and included women.

Other limitations unique to using data from the LSS and AHS cohort are important to consider in the interpretation of our results. Although we used the latest dosimetry developed for this cohort, there is a degree of uncertainty in the estimated breast dose received by each subject. This uncertainty in dose estimation is likely to be nondifferential, resulting in a bias toward the null. Given our null result, we cannot rule out the possibility that dose effect modification exists but we were unable to detect it because of this uncertainty in dose estimation. It is also possible that by including only those women with data on reproductive status, this study captures a subset of LSS and AHS members with a higher radiation dose than the cohort as a whole. This surveillance bias is a result of more complete data ascertainment from women exposed to higher radiation doses. This potential bias could explain the differences in our radiation-associated risks compared with those reported by Preston et al. (13).

The analyses of Preston et al. (13) and Land et al. (8) about breast cancer incidence in atomic bomb survivors and the recent study of Walsh et al. (18) highlight the fact that estimates of age-dependent radiation-related risks can be quite sensitive to the choice of model for the effects of birth cohort and age on background risk. However, our estimates of radiation-related risks as functions of reproductive status at exposure did not change markedly when we replaced the preferred parametric background model with a model that treated age ATB and attained age as categorical variables, suggesting that our findings are unlikely to be biased by misspecification of the background age effects.

Finally, this study had limited power to detect effect modification by reproductive status. For example, when reproductive status was included in the baseline model, the EAR among women between menarche and first birth was nearly double that of women who were premenarchal at the time of the bombing, but the CIs of these two estimates overlapped considerably. Using data from the entire female LSS and AHS cohorts, we estimated that we would have 80% power to detect an effect modification factor of ∼2.4. However, we only had data on a subset of the population used to calculate study power. Whereas we expected ∼ 1,073 breast cancer cases, we only had 641 cases that could be classified by reproductive status, reducing the study power considerably. Our negative results do not conclusively rule out the possibility of dose effect modification of smaller magnitude that might nevertheless be of biological importance.

In summary, although we found evidence of dose effect modification by reproductive status, this effect disappeared when we accounted for the heterogeneity in the baseline breast cancer risk between reproductive status groups. This suggests the possibility that radiation exerts similar carcinogenic effects on the breast regardless of its stage of differentiation pointing to the potency of radiation as a breast carcinogen.

Disclosure of Potential Conflicts of Interest

No potential conflicts of interest were disclosed.