Passive Smoking Exposure and Female Breast Cancer Mortality

Daniel Wartenberg, Eugenia E. Calle, Michael J. Thun, Clark W. Heath, Jr., Cathy Lally, Tracey Woodruff

Affiliations of authors: D. Wartenberg, Environmental and Occupational Health Sciences Institute, University of Medicine and Dentistry of New Jersey–Robert Wood Johnson Medical School, Piscataway, NJ; E. E. Calle, M. J. Thun, C. W. Heath, Jr., C. Lally, American Cancer Society, Atlanta, GA; T. Woodruff, Office of Policy, U.S. Environmental Protection Agency, Washington, DC.

Correspondence to: Daniel Wartenberg, Ph.D., Environmental and Occupational Health Sciences Institute, 170 Frelinghuysen Rd., Piscataway, NJ 08854 (e-mail: dew{at}eohsi.rutgers.edu).


    ABSTRACT
 Top
 Notes
 Abstract
 Introduction
 Subjects and Methods
 Results
 Discussion
 References
 
Background: Several studies have reported positive associations between environmental tobacco smoke (ETS) and increased risk of breast cancer. However, studies of active smoking and risk of breast cancer are equivocal and in general do not support a positive association. To try to resolve this paradox, we examined the association between breast cancer mortality and potential ETS exposure from spousal smoking in an American Cancer Society prospective study of U.S. adult women. Methods: We assessed breast cancer death rates in a cohort of 146 488 never-smoking, single-marriage women who were cancer free at enrollment in 1982. Breast cancer death rates among women whose husbands smoked were compared with those among women married to men who had never smoked. Cox proportional hazards modeling was used to control for potential risk factors other than ETS exposure. Results: After 12 years of follow-up, 669 cases of fatal breast cancer were observed in the cohort. Overall, we saw no association between exposure to ETS and death from breast cancer (rate ratio [RR] = 1.0; 95% confidence interval [CI] = 0.8–1.2). We did, however, find a small, not statistically significant increased risk of breast cancer mortality among women who were married before age 20 years to smokers (RR = 1.2; 95% CI = 0.8–1.8). Conclusions: In contrast to the results of previous studies, this study found no association between exposure to ETS and female breast cancer mortality. The results of our study are particularly compelling because of its prospective design as compared with most earlier studies, the relatively large number of exposed women with breast cancer deaths, and the reporting of exposure by the spouse rather than by proxy.



    INTRODUCTION
 Top
 Notes
 Abstract
 Introduction
 Subjects and Methods
 Results
 Discussion
 References
 
Passive exposure to environmental tobacco smoke (ETS) is an established risk factor for adult lung cancer, acute respiratory disorders (particularly in children), reduced pulmonary function, increased risk of lower respiratory infections (e.g., pneumonia and bronchitis), and, probably, ischemic heart disease (15). Five case–control studies (613) and two prospective studies (1416) have found positive associations between ETS exposure and breast cancer incidence or death; in three studies, the associations were statistically significant (Table 1Go). The interpretation of these studies is controversial, however, because active smoking is not an established risk factor for incident breast cancer (17), tobacco smoke exposure is lower from ETS than from active smoking, and most of the studies that have examined this association are relatively small and retrospective.


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Table 1. Female breast cancer risk among female nonsmokers exposed to ETS*
 
However, Wells (16,18) has proposed several reasons why ETS exposure could cause breast cancer in women, even though studies of active smoking have not revealed a positive association with breast cancer. First, previous studies (6,9,16) of active smoking have included nonsmokers exposed to ETS in the referent group, potentially obscuring an association between active smoking and breast cancer. In fact, Wells (16) hypothesizes that the lack of apparent association between breast cancer and active smoking may reflect two opposing factors: an adverse effect of ETS exposure in both active and passive smokers and a small protective effect of mainstream smoke, which interferes with estrogen, in active smokers.

A second possible explanation for the different effects of ETS and mainstream smoke on breast cancer risk is that the age at first exposure may modify the effect of exposure on breast cancer risk. Women may be at greatest risk from both ETS and mainstream smoke early in life, when the breast tissue is growing most rapidly. Failure to consider age at first exposure may obscure an association with breast cancer, particularly in the subgroup of never-smoking women exposed during adolescence (1921). Most studies of ETS and mainstream smoke do not assess this potential effect explicitly. However, Calle et al. (22) found that women who started smoking before age 16 years were 1.6 (95% confidence interval [CI] = 1.2–2.2) times more likely to die of breast cancer than nonsmokers, whereas the risk for women who started smoking at age 20 years or older was equal to that for nonsmokers. Investigators who fail to look separately at this young age group in active or passive smokers may miss the most sensitive subgroup. Some other researchers (23,24) have reported increased breast cancer risk among women who started smoking during adolescence or prior to their first pregnancy, while still others (13,25) have reported no increase in this risk. Differences in the age at first exposure in the various study populations could explain some of the differences among study results.

A third possible explanation is that genetic polymorphisms may modify the effect of smoking on breast cancer. Ambrosone et al. (26) reported that only those postmenopausal women with the slow acetylation phenotype of the polymorphic N-acetyltransferase 2 (NAT2) polymorphism showed an association between active smoking and incident breast cancer risk. Fast acetylators showed no such association. A similar evaluation of women from the Nurses' Health Study (27), in which 76% of the subjects were postmenopausal, found a nonstatistically significantly elevated breast cancer risk of 1.5 (95% CI = 0.7–3.2) among slow acetylators who currently smoked at least 15 cigarettes per day as compared with never-smoking rapid acetylators. Rapid acetylators who smoke this number of cigarettes had a smaller risk of 1.2 (95% CI = 0.5–3.3) times that of never-smoking rapid acetylators. Morabia et al. (28) reported that, among postmenopausal slow acetylators, active smokers had a breast cancer risk of 2.5 (95% CI = 1.0–6.3) times that of never smokers; among postmenopausal rapid acetylators, the relative risk was 1.3 (95% CI = 0.5–3.2). Among premenopausal women, the relative risk of breast cancer for active smokers as compared with never smokers was 0.9 (95% CI = 0.4–2.2) among slow acetylators and 1.6 (95% CI = 0.6–4.4) among rapid acetylators. However, polymorphisms of NAT2 activity have not been reported with respect to ETS exposure and menopausal status. Because most studies of active and passive smokers were not stratified on menopausal status and NAT2 genotype, we do not know whether the populations studied were similar with respect to these two factors that seem to predict breast cancer risk and whether the comparisons are interpretable meaningfully.

Finally, tobacco-specific nitrosamines and certain other carcinogens are more concentrated in ETS than in mainstream smoke (1,2,29). Therefore, it is possible that exposure to ETS confers greater risk than active smoking alone (on a per weight basis). In addition, a greater proportion of sidestream smoke than mainstream smoke is in the vapor phase than in the particulate phase (30), resulting in greater absorption of chemicals into the blood and lymph systems (31).

To further investigate the risk of breast cancer among women exposed to ETS, we examined the association of breast cancer mortality and ETS exposure in a large, prospective cohort of women enrolled in the American Cancer Society's Cancer Prevention Study II (CPS-II). This cohort has provided a wealth of data about active smoking (29,32,33) and has been used to investigate the association between ETS exposure and mortality from lung cancer (4) and coronary heart disease (5).


    SUBJECTS AND METHODS
 Top
 Notes
 Abstract
 Introduction
 Subjects and Methods
 Results
 Discussion
 References
 
Study Population

Women in this study were selected from the 676 306 female participants in CPS-II, a prospective mortality study of about 1.2 million U.S. men and women begun by the American Cancer Society in 1982 (22,34). The cohort is a convenience sample identified and enrolled by more than 77 000 volunteers in all 50 states, the District of Columbia, and Puerto Rico. Families were eligible for enrollment if at least one household member was 45 years old or older. All enrolled members were at least 30 years old. The median age of the female study participants in 1982 was 56 years; 75% of the women were between the ages of 45 and 70 years. Enrollees completed a mailed confidential questionnaire that included questions about personal identifiers: demographic characteristics; personal and family history of cancer and other diseases; and various behavioral, environmental, occupational, dietary, and (for women) reproductive exposures.

The vital status of study participants was determined from the month of enrollment through December 31, 1994, using two approaches. First, volunteers made personal inquiries in September 1984, 1986, and 1988 to determine whether enrollees were alive or deceased and to record the date and place of all deaths. Second, automated linkage using the National Death Index was used to extend follow-up through December 31, 1994 (35), and to identify deaths among 13 219 (2%) women who were lost to follow-up between 1982 and 1988. At the end of mortality follow-up in December 1994, 587 855 (86.9%) women were alive, 86 374 (12.8%) had died, and 2077 (0.3%) had had follow-up truncated on September 1, 1988, because of insufficient data for the National Death Index linkage. Death certificates or multiple causes of death codes were obtained for 98.0% of all women known to have died.

Exposure to ETS was defined in two ways. The primary analyses defined exposure to ETS as active smoking by the spouse of a never-smoking female study participant, as reported by the spouse in his questionnaire. These analyses examined current and former spousal smoking of cigarettes, cigars, and pipes and considered both amount and duration of spousal smoking. A second definition of ETS exposure was derived directly from each woman's report of the number of hours per day that she was "exposed to the smoke of others" at home, at work, and elsewhere.

Breast cancer deaths were defined as deaths through December 31, 1994, from breast cancer (International Classification of Diseases, 9th Revision, codes 174.0–174.9) (36) as the underlying cause. We excluded from all analyses women who reported prevalent cancer at baseline (except nonmelanoma skin cancer), women who had ever smoked or had an unknown smoking history, women who had no spouse in the study cohort or had been married more than once, or women who had a spouse with an unknown smoking history (Table 2Go). In addition, for the analysis of amount and duration of exposure to spousal ETS, we also excluded women whose spouse reported incomplete information regarding amount and duration of smoking and women whose age at the time of marriage was unknown (Table 2Go). For the analysis of self-reported exposure to the smoke of others, we excluded women with invalid data for these questions (Table 2Go). Invalid data were responses that were omitted or incomplete, making it impossible to determine smoking status, duration, or amount.


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Table 2. Exclusion criteria and eligible cohorts for analysis (Cancer Prevention Study II [CPS-II], United States, 1982–1994)
 
Statistical Methods

We used Cox proportional hazards modeling (37) to compute rate ratios (RRs) and to adjust for potential risk factors other than ETS. Cox models were adjusted for year of age at study entry, race (white, black, or other), years of education (<12, 12, 13–15, or >=16 years), history of breast cancer in mother or sister (yes or no), personal history of breast cysts (yes or no), age at first live birth (<20, 20–24, 25–29, or >=30 years), age at menarche (<12, 12, 13, or >=14 years), age at menopause (<45, 45–49, 50–54, or >=55 years or premenopausal), number of spontaneous abortions (0, 1, or >=2), oral contraceptive use (ever or never), use of estrogen replacement therapy (ever or never), body mass index (weight in kg/height in m2; >19.1, 19.1–21.9 22.0–27.2, 27.3–32.2, or >=32.3), alcohol intake (none, three drinks per week to fewer than one drink per day, one drink per day, or more than one drink per day), total fat consumption (estimated grams per week categorized into quintiles) (38), vegetable consumption (frequency per week categorized into quintiles), occupation (blue collar, white collar, housewife, or unknown), and spousal occupation (blue collar, white collar, househusband, or unknown).

To evaluate the possibility of overfitting the model, we calculated both the RRs adjusted for age only and the RRs adjusted for all covariates. RRs reported in the text are from multivariate models with all variables included. To evaluate potential exposure–response relationships, we included ordinal variables for packs per day smoked by the spouse (<1, 1, >1 to <2, or >=2), number of years the spouse smoked (1–10, 11–20, 21–30, or >30 years), and pack-years smoked (1–12, >12–25, >25–41, or >41 pack-years) by the spouse and used the Wald chi-square test to test for statistical significance of a linear trend (39). Pack-years were calculated by multiplying the number of packs of cigarettes smoked per day by the number of years of smoking.

To evaluate the potential effect modification, we included statistical interaction terms between age at baseline or age at marriage and the main effect (any, current, or former smoking of her spouse) and used the chi-square test to evaluate statistical significance.

To summarize the set of published passive smoking and breast cancer studies, we conducted a simple meta-analysis (40). We assessed the homogeneity of the studies using a chi-squared test. We stratified the studies by design and calculated the average risks for each stratum under a random-effects model.


    RESULTS
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 Notes
 Abstract
 Introduction
 Subjects and Methods
 Results
 Discussion
 References
 
In general, never-smoking women married to never-smoking spouses were similar in age, education, and other demographic parameters to never-smoking women married to a current or former smoker (Table 3Go). After 12 years of follow-up, 669 cases of fatal breast cancer were observed among an analytic cohort of 146 488 never-smoking married women who were cancer free at study entry and who had been married only once before enrollment (Table 2Go). Breast cancer mortality rates did not differ significantly among never-smoking women married to nonsmokers, former smokers, or current smokers (Table 4Go). Age-adjusted and multivariate-adjusted RR estimates were virtually identical. Similarly, breast cancer mortality rates did not show a statistically significant increase with the number of packs of cigarettes smoked by the spouse, the duration of spousal smoking, or the pack-years of smoking (Table 5Go). The only statistically significant association was found in women married to current smokers who had smoked for 11–20 years (RR = 2.5; 95% CI = 1.3–5.1). However, no elevations in RRs were seen in women whose currently smoking spouses had smoked for 21–30 years (RR = 1.1; 95% CI = 0.7–1.6) or >31 years (RR = 0.9; 95% CI = 0.6–1.2) (data not shown). On the other hand, two subgroups of women married to former smokers had RRs that were statistically significantly less than 1.0 (smoking duration, 11–20 years [RR = 0.5; 95% CI = 0.3–0.8]; smoking amount, >12–25 pack-years [RR = 0.6; 95% CI = 0.4–0.9]) (data not shown). Also, there was a slight elevation in risk among women married to former cigar and pipe smokers, although the increase was not statistically significant (RR = 1.3; 95% CI = 0.9–1.8) (Table 4Go).


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Table 3. Demographic characteristics of women by spousal smoking history (Cancer Prevention Study II, United States, 1982–1994)
 

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Table 4. Breast cancer mortality among single-marriage lifelong never smokers according to spousal smoking status (Cancer Prevention Study II, United States, 1982–1994)
 

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Table 5. Breast cancer mortality among single-marriage lifelong never-smoking women according to amount and duration of cigarette smoking reported by the spouse (Cancer Prevention Study II, United States, 1982–1994)
 
We also investigated two specific hypotheses on age effects on breast cancer mortality rates that had been suggested by previous studies. First, Hirayama (14), as cited in Wells (16), found that nonsmoking women married to husbands aged 40–59 years who smoked had excess breast cancer mortality relative to nonsmoking women married to husbands 40–59 years who did not smoke. Among nonsmoking women married to men over 59 years, there was no difference in breast cancer mortality between those married to smoking or nonsmoking husbands. If we assume that the age of women is roughly similar to that of the spouse, we can compare these results with those of other studies. For example, in our study, the lack of an association between spousal smoking and breast cancer mortality did not differ by age of study participants at baseline (Table 6Go). Second, Calle et al. (22), in a study of active smoking and breast cancer, reported excess risk for women who started smoking before age 20 years. Sandler et al. (7), Smith et al. (8), Lash and Aschengrau (13), and Johnson et al. (12) all reported an increased risk of breast cancer in premenopausal women exposed to ETS in childhood, although the numbers of subjects were quite small in each of these analyses. In our study, we also found a small, but not statistically significant, increase in the risk of breast cancer mortality among women married under the age of 20 years whose husbands were current smokers at enrollment into the cohort (RR = 1.2; 95% CI = 0.8–1.8) (Table 6Go).


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Table 6. Breast cancer mortality by age at baseline and age at marriage among single-marriage lifelong never smokers according to spousal smoking status (Cancer Prevention Study II, United States, 1982–1994)
 
We also examined the association between breast cancer mortality and each woman's reported exposure to ETS. No statistically significant associations were observed between reported exposures at home (RR = 1.1; 95% CI = 0.9–1.3), at work (RR = 0.8; 95% CI = 0.6–1.0), or in other places (RR = 0.9; 95% CI = 0.7–1.2) (data not shown). When reported exposures from all sources were combined and examined according to daily hours of exposure, no trend was observed (1-hour daily exposure [RR = 1.0; 95% CI = 0.8–1.2]; 2- to 4-hour exposures [RR = 1.0; 95% CI = 0.8–1.3]; 5- to 8-hour exposures [RR = 0.9; 95% CI = 0.7–1.2]; and >=9-hour exposure [RR = 0.7; 95% CI = 0.4–1.3])


    DISCUSSION
 Top
 Notes
 Abstract
 Introduction
 Subjects and Methods
 Results
 Discussion
 References
 
In this large, prospective study, we found no association between exposure to ETS from a smoking spouse (as reported by the spouse) or from other sources (as reported by the women) and a woman's risk of dying of breast cancer. Breast cancer mortality risk also was not statistically significantly associated with the amount or duration of spousal smoking, with the age of the woman at entry in the study, or with her age at marriage.

The strengths of our study are its prospective design, its size, and its use of spousal rather than proxy exposure estimates. Our analyses of spousal smoking were based on 396 breast cancer deaths among ETS-exposed, never-smoking women. Only one other study (12) included more than 150 exposed case subjects, and that was a case–control study (Table 1Go). We consider exposure to spousal ETS to be a valid indicator of potential long-term exposure to ETS. For example, exposure to spousal smoking was associated with increased mortality from lung cancer and ischemic heart disease in this cohort (4,5). Our information on spousal smoking was classified into current or former exposure and by amount and duration and was examined in women who, at the time of enrollment, had married only once. Although we observed a few elevated RRs, we believe these to be spurious because of the small size of the elevations, their inconsistency across population and exposure subgroups, and the large number of statistical comparisons considered in this study.

One limitation of our study is the reliance on breast cancer mortality rather than on incidence to identify disease. Breast cancer cases that progress to death are a small subset of all cases (41). Thus, our results could reflect the potential effects of ETS on breast cancer incidence, survival, or both. Most previous studies that have found substantial breast cancer risks associated with ETS have been studies of incident breast cancers (Table 1Go). However, it seems unlikely that the use of mortality as an end point can explain the difference between the results of this and previous studies. If, as previous studies have suggested, never-smoking women exposed to ETS truly have a doubled risk of developing breast cancer, then, for the observed mortality effect to be null, the survival rate among women with breast cancer would have to be twice as high among women who had been exposed to ETS as among women who had not been exposed to ETS. Although this survival difference may be theoretically possible, it seems highly unlikely.

A second limitation of our study lies in our assessment of exposure to ETS. We obtained complete lifetime spousal smoking histories at time of enrollment, and these reported exposures of active smoking have been shown to be highly accurate predictors of smoking-related mortality (29,32,33). However, spousal smoking was assessed only once, at baseline, and it could have been altered over the 12-year follow-up period through changes in behavior or death. A decrease in spousal smoking, which is more likely than an increase, would have biased the results toward an apparent lack of effect of ETS on breast cancer mortality. Nonetheless, we believe a woman's long-term past exposure to spousal smoking was likely to be well captured in this study and in other studies that, using the same cohort, found positive associations between exposure to ETS and lung cancer (4) and coronary heart disease (5).

By contrast, the measure of perceived ETS exposure in the home and outside the home at the time of enrollment was a crude, self-reported measure that was the only source of information on ETS exposure from household members other than the spouse and from co-workers and others outside the house. It is possible that our lack of data on past ETS exposures outside the home (or nonspousal exposures in the home) could introduce sufficient misclassification to obscure a small but true association between breast cancer and lower levels of ETS exposure. However, cumulative exposure to spousal smoking has been found to be associated with ETS exposure outside the home, and our analyses suggest that breast cancer death rates do not increase with cumulative exposure to spousal smoking (14,16). In addition, the measurement of exposure to ETS among never-smoking women in this cohort has been sensitive enough to reveal small positive associations with mortality in previous studies (4,5).

In contrast to this study, previous studies of the association of ETS exposure and female breast cancer have shown fairly consistent positive results (Table 1Go). For example, using data from Hirayama's 16-year census population-based prospective mortality follow-up study of men and women aged 40 years and above (14), the study most similar to this one, Wells (16) found that the risk of breast cancer among nonsmoking women married to smokers was 1.3 (95% CI = 0.8–2.0) times that of nonsmoking women married to nonsmokers. In the only other cohort study of never-smoking women, Jee et al. (15) found a similarly small excess breast cancer incidence among never-smoking women married to former and current smokers as compared with never-smoking women married to nonsmokers.

Five case–control studies have also yielded positive results. For example, Sandler et al. (6,7) conducted a hospital-based, case–control study of cancer among women between the ages of 15 and 59 years, somewhat younger than CPS-II women, and found a relative risk of 2.0 (95% CI = 0.9–4.3) for breast cancer among those with spousal ETS exposure as compared with women married to nonsmokers, increasing from 2.0–3.3 as the number of household members who smoked increased (7,16). A large, case–control study by Smith et al. (8), which was limited to women diagnosed with breast cancer prior to age 36 years, reported a modest risk associated with ETS exposure (odds ratio [OR] = 1.6; 95% CI = 0.8–3.1).

In an even larger population-based, case–control study in Switzerland, Morabia et al. (9) found an elevated risk of breast cancer incidence among nonactive women smokers (i.e., those who smoked <100 cigarettes in their lifetime) ever exposed to ETS (OR = 3.1; 95% CI = 1.6–6.1) as compared with those never exposed to ETS and for those women ever married to an active smoker (OR = 2.0; 95% CI = 1.1–3.7) as compared with those never married to an active smoker.

Lash and Aschengrau (13), in a case–control study of women diagnosed with breast cancer in the mid-1980s, found that passive smokers were twice as likely to be diagnosed with breast cancer as women who had not been exposed to ETS (OR = 2.0; 95% CI = 1.1–3.7). Women who were first exposed to ETS before age 12 years had an OR of 4.5 (95% CI = 1.2–16.0), and those exposed at or after age 12 years were at smaller but still elevated risk.

Finally, Johnson et al. (1012) conducted a case–control study using the Canadian National Enhanced Cancer Surveillance System, mailing a questionnaire to nearly 4000 women, more than 60% of whom responded. Premenopausal never-smoking women regularly exposed to ETS had an OR 2.3 times that of never-smoking women who were not exposed to ETS (95% CI = 1.2–4.6) (12), and postmenopausal never-smoking women regularly exposed to ETS had an OR of 1.2 (95% CI = 0.8–1.8) (11,12). However, this study had several limitations, including an extremely small referent group (14 premenopausal and 52 postmenopausal case subjects), recall bias, and response bias.

Limitations applicable to most of these studies include the small number of subjects in at least some categories, possible recall bias in case–control studies, failure to use a consistently unexposed group for the referent, uncertainty in exposure quantification, possible uncontrolled confounding, and failure to adjust for genetic susceptibility. In addition, the age of the subjects varied substantially, complicating the consideration of the possible role of menopausal status and exposure at young ages. Many of these limitations also apply to this study.

Wells (18) reported that, when combined in a simple meta-analysis, the average relative risk of the first four studies described above [Sandler et al. (6,7), Smith et al. (8), and Morabia et al. (9)] is 1.8 (95% CI = 1.4–2.4). Considering all eight studies (Table 1Go), we find the studies to be heterogeneous (P = .011). Therefore, we stratify by study design and find each group to be statistically homogeneous (cohort studies, P = .266; case–control studies, P = .218). The average risk for the case–control studies is 1.8 (95% CI = 1.4–2.5), whereas that for the cohort studies is 1.1 (95% CI = 0.9–1.4).

The disparity between the results for case–control and cohort studies is the most perplexing observation of our study. One possible explanation is that the case–control studies are biased. Typical biases found in case–control studies include recall bias and selection bias. However, the consistency of results among the case–control studies suggests that selection bias or other design-related issues are unlikely explanations for the higher risk found in the case–control studies.

The cohort studies, although prospective, had relatively short follow-up times of 16 years (14), 4 years (15), and 12 years (this study). Moreover, in each of these studies, reported smoking history reflected lifetime exposures, with inadequate recall resulting in possible misclassification. Assuming that this misclassification among healthy subjects occurred at the same rate for those exposed to ETS and those unexposed, it likely would lead to an apparent lack of effect of ETS on breast cancer mortality. However, in each of these studies, other smoking-related cancers were associated with ETS exposure among nonsmoking wives, arguing against the existence of substantial misclassification.

Another difference between the case–control and cohort studies is that all of the case–control studies used the incidence of breast cancer as their study end point, but two of three cohort studies used breast cancer mortality as their end point. Although we argued earlier in this article that we do not believe that this difference in the end point studied can explain the disparity in the results, we cannot rule it out completely.

Clarification of several issues would help us better understand the differences in the reported associations between breast cancer mortality in never-smoking women and ETS exposure. First, development of reliable, validated smoking histories would be helpful in characterizing exposures. Although we believe that changes in spousal smoking are unlikely to explain the lack of observed association between ETS exposure and breast cancer mortality among never-smoking women, particularly in the cohort studies, data to support this view would be helpful. Second, assessment of possible genetic-effect modification would help clarify whether variations in genetic makeup is a possible explanation for disparate findings. Third, the roles of adolescent exposures and menopausal status deserve further scrutiny. Although most studies adjusted for age, only a few case–control studies were stratified on either adolescent exposure or menopausal status. Of interest, stratified results for genetic interactions were stronger in postmenopausal women (26,28), whereas passive smoking exposure effects were stronger in premenopausal women (12,16).

In conclusion, given the importance of understanding the many causes of female breast cancer, the potentially high population-attributable risk of ETS exposure (even if the relative risk is modest), and the limited number of previous studies of this issue, our analyses provide an important contribution to the study of the association between exposure to ETS and female breast cancer.

Overall, the results are null. That is, we see little evidence of an association between exposure to ETS and the risk of dying of breast cancer. These results contradict results from previous studies, which were mainly case–control studies, but they are compelling in that they come from a large prospective cohort study. We did find data that suggest that women exposed to ETS under age 20 years may be at increased risk of dying of breast cancer, however, and this observation warrants follow-up.


    NOTES
 
These views represent those of the authors and do not necessarily represent those of the U.S. Environmental Protection Agency.

Supported by U.S. Environmental Protection Agency grant 992097-1 and by Public Health Service grant ES0502209 from the National Institute of Environmental Health Sciences, National Institutes of Health, Department of Health and Human Services.

We thank Alfredo Morabia for providing helpful comments on the manuscript.


    REFERENCES
 Top
 Notes
 Abstract
 Introduction
 Subjects and Methods
 Results
 Discussion
 References
 

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Manuscript received January 4, 2000; revised August 14, 2000; accepted August 16, 2000.


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