P for trend=0.0001 and 0.01, respectively), but not of ER − PR − tumours (P for trend=0.27 and 0.85), whereas duration of breastfeeding was associated with lower relative risk of both receptor-positive (P for trend=0.0002) and receptor-negative tumours (P=0.0004). Our results were consistent across subgroups of women based on age and ethnicity. We found few significant differences by histologic subtype, although the strongest protective effect of multiparity was seen for mixed ductolobular tumours. Our results indicate that parity and age at first birth are associated with reduced risk of receptor-positive tumours only, while lactation is associated with reduced risk of both receptor-positive and -negative tumours. This suggests that parity and lactation act through different mechanisms. This study also suggests that reproductive factors have similar protective effects on breast tumours of lobular and ductal origin.Reproductive factors are associated with reduced risk of breast cancer, but less is known about whether there is differential protection against subtypes of breast cancer. Assuming reproductive factors act through hormonal mechanisms they should protect predominantly against cancers expressing oestrogen (ER) and progesterone (PR) receptors. We examined the effect of reproductive factors on subgroups of tumours defined by hormone receptor status as well as histology using data from the NIHCD Women's Contraceptive and Reproductive Experiences (CARE) Study, a multicenter case–control study of breast cancer. We estimated odds ratios (ORs) and 95% confidence intervals (CIs) as measures of relative risk using multivariate unconditional logistic regression methods. Multiparity and early age at first birth were associated with reduced relative risk of ER + PR + tumours ( The epidemiologic evidence that reproductive factors such as number of full-term pregnancies, age at first full-term pregnancy and duration of breastfeeding are strongly predictive of breast cancer risk is substantial . We receBoth multiparity and age at first pregnancy may exert their protective effects on breast cancer through hormonal mechanisms  and nonhThe histology of the tumour may complicate this picture even further. It has been hypothesised that lobular cancers may be more strongly associated with certain hormonal risk factors such as postmenopausal hormone therapy use , while if they act through a nonhormonal mechanism, they will be associated with hormone receptor-positive and -negative tumours to an equal extent.We also evaluate the associations between reproductive factors and development of tumours of different histologies in all women combined as well as by ERPR status.The Women's CARE Study is a population-based case–control study of breast cancer designed to address the role of reproductive, contraceptive and lifestyle factors on the breast cancer risk of White and African-American women ages 35–64 years in five US regions . The design and methods of this study have previously been described in detail  cancer registries in Atlanta, Seattle, Detroit and Los Angeles, or from hospitals in Philadelphia. We restricted eligibility to US-born English-speaking women. We oversampled younger case patients and African-American case patients to maximise their numbers in the study, and we randomly sampled older White case patients to achieve approximately equal numbers of case patients in each 5-year age group. Of the 5982 eligible case patients, we interviewed 4575 (76.5%), including 2953 White patients and 1622 African-Americans.in situ breast cancer. Control subjects were frequency matched to case patients by study centre, race and 5-year age group. We successfully screened 82% of residential households called. Of the 5956 eligible women selected as control subjects, we interviewed 4682 (78.6%), including 3021 White subjects and 1661 African Americans.Control subjects were randomly selected from a pool of eligible women identified through random digit dialling (RDD) identified between July 1994 and June 1998. They were US-born English-speaking women who had never been diagnosed with invasive or Risk factor information was obtained from an in-person interview using a structured questionnaire that included questions on demographics, reproductive history (including breastfeeding), medical history including use of exogenous hormones, family history of cancer and other lifestyle factors. Information was recorded up to the date of diagnosis (month and year) for case patients or the date of initial household contact by RDD for control subjects. All cases and controls were interviewed between September 1994 and December 1998. Informed consent was obtained from all participants. Study protocols were approved by the Institutional Review Board at each participating centre.Oestrogen receptor and PR status of case patients was recorded from the pathology record when available at each study centre. All SEER registries routinely collect this information for breast cancer, specifically they collect the laboratory results recorded in the medical records at the time of diagnosis. Overall, 3969 (86.8%) of case patients had ER status data available. Among those missing ER status, the pathology record indicated that the test had not been carried out for 153 cases, it had been ordered but was not available for 171 cases and the information was missing for 252 cases. An additional 30 cases had ER status listed as borderline and were also excluded from these analyses. We had PR information on 3795 women (83.0%). An additional 179 were ordered but results were not available, 238 were not carried out and no information was available on 319. The PR status of 44 was borderline and these were excluded from the analyses. Both ER and PR status were available on 3771 cases, or 82.4%. The percentage of participants with both ER and PR status available by site were: Atlanta 91.5%, Detroit 70.3%, Los Angeles 79.0%, Philadelphia 72.3% and Seattle 93.5%.We conducted separate analyses by histology for 3455 ductal tumours , 274 lobular tumours (ICD-O morphology code 8520) and 261 mixed ductolobular tumours (ICD-O morphology code 8522). An additional 577 tumours of other histologies were not included in these analyses.vs never pregnant), parity ), number of pregnancies, number of full-term pregnancies, age at first full-term pregnancy, years since last full-term pregnancy and duration of breastfeeding.We evaluated the following reproductive variables: gravidity (ever pregnant We excluded 21 women (seven cases and 14 controls) from the analyses because of unknown information on one or more pregnancy or breastfeeding variable These exclusions left us with 4668 controls and 3764 cases for the ERPR analyses. For the histology analyses, we excluded one additional case with missing parity and ERPR information. Thus, we ended up with 3990 cases for the histology analyses. A total of 11 women who had unknown ages at menarche were assigned the median age of 12 years and retained in all analyses.a priori. The following variables were included in all models: age , race (White and African-American), family history of breast cancer  family history, unknown or adopted), age at menarche , study centre (five sites) and education . Where appropriate, we also adjusted for age at first full-term pregnancy  and number of full-term pregnancies . For analyses of all women combined, we used never pregnant women as the reference group. For the analyses of age at first birth and total breastfeeding, we only included parous women. The reference group for total breastfeeding was parous women who had never breastfed. All adjustment variables were included as categorical variables in the models. We report the P-values from trend tests  from the case–control analyses in the tables. All P-values reported are two-sided.For each stratum of ERPR status, we estimated odds ratios (ORs) and 95% confidence intervals (CIs) as measures of relative risk using unconditional logistic regression methods , while cTo determine whether the reproductive factors were associated equally with breast cancer of different receptor statuses, we excluded the controls, and defined ER + PR + cases as ‘cases’ and ER − PR − cases as ‘controls’. We determined the ORs of being a receptor-positive case associated with the various reproductive factors, and estimated the tests for association or trend. A statistically significant test of association or trend (likelihood ratio test) from the case–case analysis suggested that the effect of the exposure factor differed by receptor status. A similar procedure was used for comparing ER + PR − cases and ER − PR + cases with ER − PR − cases, as well as for histology status, where we evaluated whether either lobular or ductolobular cancers differed from ductal tumours.All analyses were performed using EPILOG  or the SAS statistical package .Study protocols were approved by the Institutional Review Boards at each participating centre.χ2P=0.01), somewhat later age at first birth (P=0.0001), and breastfed longer (P=0.03) than those without data on ER and PR , we focus our presentation on the differences between the ER + PR + and the ER − PR − cancers. Ever having had a full-term pregnancy was associated with a reduced OR for ER + PR + tumours , while no such effect was observed for ER − PR − tumours . Being multiparous was associated with reduced ORs of ER + PR + tumours, but not of ER − PR − tumours (P comparing cases by ERPR status=0.00006), with five or more full-term pregnancies conferring an estimated 58% reduction in risk of ER + PR + tumours.Cases with data on ER and PR had fewer children (R and PR . Table 2P=0.07).Among parous women, older ages at first birth were associated with modest increases in relative risk estimates of ER + PR + tumours, but not ER − PR − tumours, although the test comparing receptor status did not achieve statistical significance (P comparing cases by ERPR status=0.71).On the other hand, the effects of breastfeeding did not differ by receptor status. Among parous women, the relative risk estimates for both ER + PR + and ER − PR − tumours were statistically significantly reduced among those who had breastfed compared to those who had not. The effect of total duration of breastfeeding was remarkably similar for both ER + PR + and ER − PR − tumours. Compared to women who never breastfed, women who breastfed 24 months or longer were at an estimated 31 and 35% lower relative risk of ER + PR + and ER − PR − tumours, respectively (P for trend=0.00001) and for ductolobular tumours (P for trend=0.001). There was no significant protective trend for lobular tumours (P=0.19), but having had four or more pregnancies was associated with an OR of 0.56, similar to the OR of 0.62 for ductal tumours and 0.43 for ductolobular tumours. The tests comparing tumours of other histologies with ductal tumours were not statistically significant .In P for trend=0.02); this was not observed for ductal (P for trend=0.32) or ductolobular tumours (P for trend=0.92). The test for comparing lobular and ductal cancer cases did not reach statistical significance (P=0.11).Women with later age at first full-term pregnancy were at higher relative risk of lobular tumours than women with a first full-term pregnancy before age 20 . We observed no statistically significant differences in the relative risks estimates for 12 months of breastfeeding between tumours of the different histologic subtypes (results not shown).When we did further analyses of tumours of different histologies across different age and ethnic groups, the results were consistent with those reported above (results not shown). When analyses were limited to White women above age 50 with ER + PR + tumours, the only subgroup defined by age, race and receptor status for which there were sufficient number of subjects to analyse histology, the difference between ductolobular and ductal tumours reached marginal statistical significance  and OR for ductolobular tumours 0.78 (95% CI 0.65–0.93), When analyses were restricted to women who had never used hormone therapy, the results were very similar, with no differences between the relative risk estimates of the different histologic types for either full-term pregnancy or breastfeeding.Our recent evaluation of reproductive factors from the CARE study  ER − PR − cases (Previous studies of the effects of parity or age at first full-term pregnancy (or age at first birth) on tumours of different receptor statuses have been mixed. Two studies found similar effects of parity on receptor-positive and -negative tumours   is remarkably similar to that reported by a prior study conducted within the SEER registries . In thatIn conclusion, high parity and early age at first birth were associated with a reduction in risk only for ER + PR + tumours. Breastfeeding was associated with a reduction in risk for both ER + PR + and ER − PR − tumours. Combined with previous research, this suggests that parity and age at first birth act through different mechanisms than breastfeeding. All reproductive factors showed similar associations with both ductal, ductolobular and lobular tumours, suggesting that these tumours have similar aetiologies.


Time-related effects of a pregnancy on breast cancer risk were examined in a population-based prospective study of 802,457 parous Norwegian women aged 20-56 years. The mean follow-up time was 16.4 years. A total of 4787 women were diagnosed with breast cancer. We observed a short-term increase in risk of breast cancer after a full-term pregnancy, with a maximum 3-4 years after delivery, followed by a long-lasting decrease in risk. The maximum risk was about twice the risk for women whose last delivery was 20 or more years previously . Compared with nulliparous women, those with one or two children were at higher risk in the first decade after the last pregnancy, whereas those with three or more children were at lower risk in most categories of time since the last birth. The positive association between breast cancer risk and age at last birth was markedly reduced after adjustment for time since last birth. We conclude that there is a non-linear relationship between breast cancer incidence and time since last birth. Part of the relation with age at last birth may be attributed to the association with time since last birth.


Active smoking has little or no effect on breast cancer risk but some investigators have suggested that passive smoking and its interaction with active smoking may be associated with an increased risk. In a population based case–control study of breast cancer in women aged 36–45 years at diagnosis, information on active smoking, passive smoking in the home, and other factors, was collected at interview from 639 cases and 640 controls. Women were categorised jointly by their active and passive smoking exposure. Among never smoking controls, women who also reported no passive smoking exposure were significantly more likely to be nulliparous and to be recent users of oral contraceptives. Among those never exposed to passive smoking, there was no significant association between active smoking and breast cancer, relative risk (RR) of 1.12  0.72–1.73) for past smokers and RR of 1.19 (95% CI 0.72–1.95) for current smokers, nor was there an association with age started, duration or intensity of active smoking. Compared with women who were never active nor passive smokers, there was no significant association between passive smoking in the home and breast cancer risk in never smokers, RR of 0.89 (95% CI 0.64–1.25), in past smokers, RR of 1.09 (95% CI 0.75–1.56), or in current smokers, RR of 0.93 (95% CI 0.67–1.30). There was no trend with increasing duration of passive smoking and there was no heterogeneity among any of the subgroups examined. In this study, there was no evidence of an association between either active smoking or passive smoking in the home and risk of breast cancer. Active smoking has little or no effect on a woman's risk of developing breast cancer , althougWe report here on the relationship between exposure to active smoking and passive smoking within the home and breast cancer risk in women aged 36–45 years using data from a UK population based case–control study which was designed to investigate the relationship between oral contraceptive use and breast cancer risk, and also collected data on a range of lifestyle factors including active and passive smoking history.This is a population based case–control study of breast cancer that recruited women who were residents of the Thames, Oxford and Yorkshire regions, aged 36–45 years who were diagnosed between July 1987 and February 1990. The study was restricted to white women with no previous malignancy (except non-melanoma skin cancer), severe mental handicap or significant illness.Cases were women with an incident invasive breast cancer, identified chiefly through the Regional Cancer Registries, supplemented by hospital discharge records from Hospital Activity Analysis registers and patient lists kept at major treatment centres. These latter two methods were used to help minimise the delay between diagnosis and interview. All cases were confirmed by cancer registration. Permission from hospital consultants and general practitioners was sought before the cases’ identities were disclosed from the Cancer Registry and before the women were contacted. Breast cancer patients’ case-notes were reviewed, confirming the diagnosis in 97% of the cases. Out of 838 eligible cases, 644 (77%) were interviewed. Reasons for nonparticipation were consultants or general practitioners’ refusal (6%), woman's own refusal (3%), death (11%) and relocation from the study area (3%).For every participating case, one age-matched control was randomly selected from patients registered with the case's general practitioner. If a chosen control could not be interviewed, a further age-matched control was selected. Of the 644 first eligible controls, 588 (91%) were interviewed. Reasons for nonparticipation of the primarily selected controls were general practitioners’ refusal (3%) or woman's own refusal (6%).Study participants were interviewed in their homes, using a standardised structured questionnaire. Each case–control pair was interviewed by the same, trained, interviewer. The mean time between diagnosis and interview was approximately 2 years (range 1–5 years). Women were asked whether they had ever smoked, and if so, how old they were when they started. For each subsequent year, women were asked to report the number of months of smoking and the average number of cigarettes smoked per day . In addition, the participants were asked to report for each year from the age of 16 years if they were married or living with a boyfriend, and if yes how many cigarettes per day he smoked at home . In addition, information was sought from each woman about her menstrual and reproductive history and other risk factors for breast cancer.Women were defined as current smokers if they reported being a smoker in the year prior to interview, and as past smokers if they had stopped smoking for at least 1 year prior to interview. Women were categorised as being passive smokers if they reported living for at least 1 year with a partner who smoked at home. Among women who were passive smokers, the total number of years of exposure and the cigarette-years of exposure were calculated, the latter as the number of years of exposure multiplied by the number of cigarettes smoked per day in the home.n=14), bilateral oophorectomy (n=6) or were users of hormone replacement therapy (n=73). One control who did not have information on active smoking exposure along with three controls and five cases who did not have information on passive smoking exposure were excluded leaving 639 cases and 640 controls for analysis.Women were defined as peri/postmenopausal if they reported having had a natural menopause  and 95% confidence intervals (CIs) were calculated using conditional logistic regression with stratification by age at diagnosis/pseudodiagnosis (single year) and region of recruitment. All estimates are presented stratified by age and region and additionally adjusted for socioeconomic status of the woman , nonmanual (SC3), manual/not employed (SC4/SC5/other)), alcohol consumption , BMI , parity and age at first giving birth , use of oral contraceptives , family history of breast cancer , age at menarche (<13 and 13+ years) and menopausal status . These adjustments were included on the basis of a priori evidence of their effects on breast cancer risk irrespective of their significance in this study. The effect of passive smoking exposure was examined in a number of subgroups known to be related to breast cancer risk – menopausal status, alcohol consumption, oral contraceptive use, family history of breast cancer, parity and age at first birth, socioeconomic status, BMI and age at menarche. In these analyses, RRs were estimated from conditional logistic regression models stratified by age and region and adjusted for parity  and oral contraceptive use  where appropriate. Heterogeneities between estimates of RRs were assessed using χ2 tests.Differences in characteristics were assessed using either P-values less than 0.05 were regarded as statistically significant.All analyses were performed in Stata version 8.1 . All statistical tests were two sided, and Characteristics of the cases and controls included in these analyses are presented in Among participants, 300 (23%) reported being a past smoker and 372 (29%) reported being a current smoker, and a higher proportion of ever smokers also reported exposure to passive smoking in the home . Similar rates of exposure to passive smoking in the home by case and control status were reported for never smokers  and ever smokers . Among smokers, the average number of cigarettes per day, duration of smoking and age started smoking were similar between cases and controls. The duration of exposure to passive smoking was similar between cases and controls for never, past and current smokers.−1 compared with 4.4 g day−1 in women not exposed. However, there were significant differences between exposure to passive smoking and both parity and oral contraceptive use  for past smokers and an RR of 1.04 (95% CI 0.79–1.36) for current smokers compared with never smokers . SimilarAmong the never smokers, there was no significant heterogeneity between risk of breast cancer and exposure to passive smoking for a range of participant characteristics . The assIn this study, neither being an active smoker nor exposure to passive smoking in the home from a partner was associated with breast cancer risk among women aged 36–45 years. There was no association between smoking intensity, duration or age started smoking and breast cancer risk, neither was there a trend with increasing duration of exposure of passive smoking. Adjustment for established breast cancer risk factors made little difference to risk estimates, and the lack of association between passive smoking and breast cancer risk was consistent among a number of subgroups including menopausal status, alcohol consumption, oral contraceptive use, family history of breast cancer, parity and age at first giving birth, socioeconomic status, BMI and age at menarche.The lack of an association between active smoking and breast cancer risk in the current study is consistent with previous reviews and a collaborative reanalysis of worldwide data . ContrarThere remains considerable controversy over the relationship between passive smoking and breast cancer since the results from previous studies are inconclusive; some suggested an increase in risk with exposure to passive tobacco smoke  and otheThe two reviews that suggested there was an association between passive smoking and breast cancer also highlighted that the relationship appeared stronger for premenopausal compared with postmenopausal breast cancer  and controls. However, at the time this study was carried out, there was less public concern over the dangers of passive smoking and therefore less chance that such retrospective reporting bias would be an issue. Since this study was performed, the prevalence of smoking among women has fallen consistently year on year and the likelihood of exposure to passive smoke both in the workplace and the home has also fallen. However, neither of these changes will have any impact on the generalisability of the results of this study, since they only affect the prevalence of the exposure variable. It is possible that this study failed to recruit heavy persistent smokers who may have died prior to being eligible for this study; however, since it recruited people before the age of 45 years where tobacco-related death rates are low, this is unlikely to be a major source of bias. A strength of the study is that the data were collected during in-person interviews by a small number of trained nurses and is complete. Furthermore, the study obtained high-quality measurements of both the smoking and passive smoking exposure, with a full history of dates starting smoking, stopping smoking and numbers of cigarettes per day being recorded whenever significant changes occurred.In summary, the current study finds no significant associations between either active smoking or passive smoking in the home and risk of breast cancer in women aged 36–45 years.


The fertility pattern influenced the breast cancer risk in this irradiated population in a similar way to that observed in other studies. SIR at dose = 0 was highest, 2.31, among postmenopausal nulliparous women . SIR at dose = 0 was lowest in pre- or postmenopausal women with a first childbirth before 25 years of age; 0.89 (0.71–1.09) and 0.88 (0.58–1.25) respectively. Thus, in addition to the dose–effect response in the cohort, part of the breast cancer excess could be explained by a different fertility pattern. The estimates of ERR/Gy for the various categories of age at first childbirth, number of children, menopausal status and ovarian dose were very similar, contradicting any interaction effects on the scale of relative risk. © 2001 Cancer Research Campaign http://www.bjcancer.comExposure to ionizing radiation is a known risk factor for breast cancer and the fertility pattern is a recognized modifier of breast cancer risk. The aim of this study was to elucidate the interaction between these 2 factors. This study is based on a Swedish cohort of 17 202 women who had been irradiated for skin haemangiomas in infancy between 1920 and 1965. The mean age at treatment was 6 months and the median breast dose was 0.05 Gy (range 0–35.8 Gy). Follow-up information on vital status, parity, age at first childbirth and breast cancer incidence was retrieved through record linkage with national population registers for the period 1958–1995. Analyses of excess relative risk (ERR) models were performed using Poisson regression methods. In this cohort, the fertility pattern differed from that in the Swedish population, with significantly fewer childbirths overall and before 25 years of age but more childbirth after that age. There were 307 breast cancers in the cohort and the standardized incidence ratio (SIR) was 1.22 (95% CI 1.09–1.36). A linear dose–response model with stratification for fertility pattern and menopausal status resulted in the best fit of the data. ERR/Gy was 0.33 (95% CI 0.17–0.53). In absolute terms this means an excess of 2.1 and 5.4 cases per Gy per 10


Examination of parity, age at menarche and at menopause by HER2 status in a large series of breast carcinomas showed a statistically significant increased-frequency of HER2-positive tumours in lower risk subgroups. The findings suggest a difference in the protective role of hormone-related risk factors between HER2-positive and -negative tumours. Clinical and molecular biology findings have led to the identification of tumour overexpressing HER2 (HER2+) as a distinct subset (comprising 20–25%) of breast carcinomas  and age at menarche (1692 cases) and at menopause (1050 cases). The subgroup with 0 or 1 child, and the subgroups of <12 or <45, respectively, for age of menarche and age of menopause were considered as control group (OR=1). The expected proportions (EP) of HER2-positivity were calculated assuming that HER2-positive tumours are not affected by hormonal risk factors and using the mean odds ratios (OR) reported for parity, ages at menarche and menopause  of HER2-positive tumours, according to the parity, suggests that this factor protects only the HER2-negative tumours. Accordingly, assuming protection only for the HER2-negative tumour subset and using the ORs as described in Patients and Methods for parity, the expected proportions of HER2-positivity would be of 20.9, 22.0 and 27.9% in the respective groups, that is, quite similar to the observed frequencies (The first step in the analysis considered the effect of parity on HER2-positive breast cancer risk. The proportion of HER2-positive tumours was found to vary from 20.9% in the group with zero or one child, to 23.8% in the group with two or three children, up to 30.5% for women with more than three children (The two other hormonal risk factors for breast carcinomas recorded in our database, that is, ages at menarche and menopause, also appear to have a protective impact only on HER2-negative tumours. In fact, an increased HER2-positive tumour proportion was observed with increased age at menarche, which is associated with decreased risk, and a decrease in HER2-positivity was observed with increased age at menopause, which is associated with increased risk . Again uAltogether our analyses suggest that the three hormone-related risk factors analysed seem to protect only HER2-negative tumour subset. These results may be interpreted also as opposite to a promoting effect of hormone-related risk factors for the HER2-positive subgroup. Indeed, some clinical data (To date, the majority of risk factors for breast carcinomas identified are related to hormones. Identification of other risk factors, specific for the onset of HER2-positive tumours might be hampered by a clouding effect of HER2-negative tumours, which represent the great majority of breast carcinomas. Indeed, as long as HER2-positive tumours, which represent only 25% of cases and therefore of little weight among all breast carcinomas, are analysed together with the negative ones, it will be difficult to sort out risk factors for HER2-positive tumours.In conclusion, our study suggests that HER2-negative and -positive tumour subsets are influenced differently by breast carcinoma hormone-related risk factors. To confirm these data, a cohort study specifically addressing the role of hormone risk-factors for breast cancer with respect to HER2 status is going on.


Diabetes during pregnancy is related to enhanced fetal growth, which has been associated with increased breast cancer risk. Whether daughters of mothers with a diagnosis of diabetes have an increased risk of breast cancer is not known.We performed a retrospective cohort study of daughters of mothers with diabetes by linkage of the Swedish Multigeneration, Cause-of-Death and Patient Register between 1952 and 2005. Breast cancer cases were ascertained by linkage with the Swedish Cancer Register between 1958 and 2005. Standardized incidence ratios (SIRs) of breast cancer were calculated assuming a Poisson distribution for the observed cases.We identified 291,360 daughters of mothers with a diagnosis of diabetes before or after birth between 1952 and 2005. Among the daughters, 7,956 cases of breast cancer were diagnosed between 1964 and 2005. The total time of follow-up was 12,173,821 person years. The expected number of breast cancer cases was 9,204, resulting in an SIR of 0.86 . The decrease in risk associated with maternal diabetes was stronger for premenopausal (< 55 years of age) than postmenopausal (≥ 55 years of age) breast cancer . Among daughters of mothers with diabetes, a history of breast cancer in the mother increased the risk of breast cancer in the daughter .Daughters of mothers with a lifetime history of diabetes were at a decreased risk of breast cancer. The strongest negative association was found among premenopausal breast cancer. In 1990, Trichopoulos suggested that breast cancer risk was influenced by the fetal environment, particularly variations in hormone concentrations  reports The study base from which our study population was drawn is the total female Swedish population from 1932 and onwards. The unique national registration number (NRN), assigned at birth or immigration, allowed linkage to the following population-based nationwide registers: the Multigeneration Register, the Patient Register, the Cancer Register, the Register of Causes of Death, the Population Register, and the Register of Population Changes  of breast cancer. SIRs were calculated by dividing the observed number of breast cancers with that expected, based on sex-, age-, and calendar period-specific incidence rates in the general population of Sweden. Confidence intervals of 95% (95% CI) were calculated assuming a Poisson distribution for the observed number of cases . The datThrough linkage among the Multigeneration Register, the Patient Register and the Cause of Death Register, we identified 291,360 daughters of mothers with a diagnosis of diabetes mellitus or gestational diabetes before or after birth from 1932 onwards. Among the daughters there were 7,956 cases of breast cancer between 1958 and 2005. The total time of follow-up was 12,173,821 person years. The expected number of breast cancer cases was 9,204, with a total SIR of 0.86 . The SIRs for breast cancer by the mother's age at diagnosis of diabetes are listed in Table In this retrospective cohort study on close to 8,000 breast cancer cases we observed a significantly decreased risk of breast cancer among daughters of mothers with a lifetime history of diabetes. The strongest negative association was found for premenopausal breast cancer, with a 17% reduction in risk. Hence the findings from our study are in contradiction to our hypothesis of an elevated risk of breast cancer among daughters of mothers with diabetes.Previous studies have found that increased fetal growth measured by birth weight and height is positively associated with breast cancer risk. In a meta-analysis including studies published between 1980 and 2007, Xue et al. report a small but significant 15% increased risk of breast cancer for increased birthweight and an almost 30% risk increase for high birth length [We did not have information on gestational age and birth weight of the daughters in the study, and therefore we could not explore whether the inverse association between maternal diabetes and breast cancer risk among the daughters was influenced by these perinatal factors. Another limitation of the study is that we were not able to distinguish between type 1, type 2, and gestational diabetes among mothers. Type 2 diabetes is by far the most common diabetes type, and the underlying biologic mechanisms affecting maternal glucose metabolism are similar for all types of diabetes. We were able to identify diabetes in the mother starting in 1965. Hence, diagnosis of diabetes in the mothers could be after the birth of the daughters being followed for breast cancer. We do not know whether the mother had previously been diagnosed with diabetes, in particular, prior to her pregnancy with the daughter, or whether the hospitalization was related to newly diagnosed diabetes; it is possible that the mother developed diabetes after the index daughter was born. Since diabetes develops over a considerable time period, however, the mother is likely to have been affected by insulin resistance and hyperinsulinemia much earlier and thus also during her pregnancy with the daughter [Fetal weight is strongly correlated with maternal weight in early pregnancy and weight gain during pregnancy . OverweiPrevious studies suggest a modest increase in risk of breast cancer among women with type 2 diabetes and gestational diabetes . In the Breast cancer incidence is higher in women with high socioeconomic status. In a recent UK study, the relative risk of breast cancer among the most deprived women was 16 percent lower than in the least deprived . AffluenBy comparing breast cancer incidence in daughters of mothers with diabetes to the general population, we probably underestimate the true association because diabetes often goes undiagnosed for several years. The prevalence of diabetes in Sweden has not increased between 1986 and 1999, and would therefore not influence the results . Only feIn conclusion we found that daughters of mothers with a lifetime history of diabetes were at a decreased risk of breast cancer. The strongest negative association was found among premenopausal breast cancer.BMI: body-mass index; CI: confidence interval; EPIC: European Prospective Investigation into Cancer and Nutrition; HRT: hormone replacement therapy; ICD: International Classification of Disease; IGF: insulin-like growth factor; LGA: large-for-gestational-age; NRN: national registration number; SHBG: sex hormone-binding globulin; SIR: standardized incidence ratio.The authors declare that they have no competing interests.OS participated in the study design and interpretation of the results and drafted the manuscript. FG performed the statistical analyses. AE participated in the design of the study and interpretation of the results. KM participated in the design, interpretation of results and coordination, and helped to draft the manuscript. All authors read and approved the final manuscript.


The present study set out to evaluate perceptions of risk, psychological morbidity and health behaviours in women with a family history of breast cancer who have attended genetic counselling and determine how these differ from general population risk women. Data were collected from 62 genetic counselees (cases) attending the Royal Marsden and Mayday University Hospital genetic counselling services and 62 matched GP attenders (controls). Levels of general psychological morbidity were found to be similar between cases and controls; however, cases reported significantly higher breast cancer-specific distress despite clinic attendance . Although cases perceived themselves to be more susceptible to breast cancer, many women failed correctly to recall risk figures provided by the clinic; 66% could not accurately recall their own lifetime chance. Clinics appeared to have a positive impact on preventive behaviours and cases tended to engage more regularly in breast self-examination , although few differences were found between groups in terms of health beliefs. We conclude that counselees and GP controls showed considerable similarities on many of the outcome measures, and risk of breast cancer was not predictive of greater psychological morbidity; although cases were more vulnerable to cancer-specific distress. Despite genetic counselling, many cases continued to perceive their risk of breast cancer inaccurately.


P=0.15). Diethylstilbestrol did not increase mortality from gynaecologic cancers. In summary, diethylstilbestrol was associated with a slight but significant increase in all-cause mortality, but was not significantly associated with overall cancer or gynaecological cancer mortality. The association with breast cancer mortality was more evident in trial participants, who received high DES doses.We used Cox regression analyses to assess mortality outcomes in a combined cohort of 7675 women who received diethylstilbestrol (DES) through clinical trial participation or prenatal care. In the combined cohort, the RR for DES in relation to all-cause mortality was 1.06 (95% CI=0.98–1.16), and 1.11 (95% CI=1.02–1.21) after adjusting for covariates and omitting breast cancer deaths. The RR was 1.07 (95% CI=0.94–1.23) for overall cancer mortality, and remained similar after adjusting for covariates and omitting breast cancer deaths. The RR was 1.27 (95% CI=0.96–1.69) for DES and breast cancer, and 1.38 (95% CI=1.03–1.85) after covariate adjustment. The RR was 1.82 in trial participants and 1.12 in the prenatal care cohort, but the DES–cohort interaction was not significant ( Although clinical trials conducted in the 1950s showed DES was not effective , use conarcinoma , and 4 marcinoma  were givMost studies, but not all  and the Dieckmann Study, both of which assessed the long-term health consequences of DES exposure during pregnancy. The WHS enrolled DES-exposed and unexposed women who were ascertained through a retrospective review of obstetrics records for the period 1940–1960 at the Mayo Clinic in Rochester, MN; Mary Hitchcock Memorial Hospital in Hanover, NH; a pregnancy clinic at the Boston Lying-In Hospital, in Boston, MA; and a private obstetrics practice in Portland, ME  was documented in the medical record, the occurrence of cancer was confirmed by medical record, and cause of death was verified by death certificate.Follow-up of the combined cohort was implemented by the National Cancer Institute (NCI) in 1992, at which time intense tracing efforts were used to locate the women who previously participated in the WHS and Dieckmann studies. In conjunction with the 1994 data collection phase, we obtained either a questionnaire (including proxy questionnaires) or a death certificate for 6495 (84%) of the 7758 women initially ascertained for the WHS and Dieckmann studies. Deaths occurring through the 1994 data collection phase were ascertained by death certificate; subsequently, mortality and cause of death were ascertained through the National Death Index. For a subset of the WHS women , the underlying cause of death was determined through the Social Security Death Index. Follow-up through both sources continued through 1 January 2000.For the present analysis, the outcomes were all-cause mortality, cause-specific mortality, including overall cancer death, and cause-specific cancer death. Diseases were classified by trained nosologists according to the International Classification of Diseases (ICD), using both ICD-9  and ICD-The analyses are based on 7675 of the 7758 women from the original cohorts (83 women were omitted owing to lack of information on study entry date). Covariate information was available for 7062 (92%) women, based predominantly on the 1994 questionnaire responses, or on earlier questionnaire responses when necessary. Cox proportional hazards models , with ti−2)  were considered when available and appropriate for specific mortality outcomes. Breast cancer models contained terms for family history of breast cancer, BMI, age at first pregnancy , and parity ; ovarian cancer analyses were adjusted for parity. We found no evidence of confounding by these variables or by cohort (RR estimates changed less than 10%); consequently, the RR shown in the tables are adjusted for birth year and study entry year. Missing covariate data were included in models by use of indicator variables.Year of birth and study entry year were covariates in all models. Additional covariates, including education , smoking , and body mass index (BMI)  . When brThe combined grouping of cerebrocardiovascular disease accounted for about one-third (33.8%) of all deaths, but there was little evidence that DES increased overall cerebrocardiovascular mortality or mortality from cardiovascular, cerebrovascular, or other vascular conditions.The slightly elevated RR for DES in relation to mortality owing to infectious disease, neuropsychiatric conditions, Alzheimer's disease, digestive diseases, and diabetes were compatible with chance. A moderately strong and statistically significant association was observed between DES exposure and deaths owing to kidney/urinary tract disease , and the RR was 2.57 (95% CI=1.23–5.39) after further adjustment for smoking. Mortality from violence/accident was similar for the exposed and unexposed women. There was little evidence of an association between DES and noncancer mortality classified as ‘other’; RR=0.90 (95% CI=0.62–1.29). Inverse associations were observed for pulmonary and liver disease mortality in relation to DES exposure, but were consistent with chance, and the results were similar after adjustment for smoking (data not shown).P=0.11). After further adjustment for BMI and smoking, and with breast cancer deaths omitted, the RR for overall cancer mortality was 1.09 (95% CI=0.94–1.28) in the combined cohort, 1.06 (95% CI=0.90–1.26) in WHS, and 1.35 (95% CI=0.92–1.97) in the Dieckmann cohort.Overall, 37.7% of deaths were owing to cancer . In the P=0.15). The association with breast cancer mortality did not change over time; the RR was 1.31 and 1.25, respectively, for less than 30 years, and 30 or more years since exposure.In the combined cohort, the RR was 1.27 (95% CI=0.96–1.69) for the association between DES and breast cancer death; after further adjustment for family history of breast cancer, BMI, age at first pregnancy, and parity, the RR was 1.38 (95% CI=1.03–1.85). The RR was 1.12 (95% CI=0.80–1.55) in the WHS cohort and 1.82 (95%=1.04–3.18) in the Dieckmann cohort, but the interaction between DES and cohort in relation to breast cancer mortality was not statistically significant ; the RR was essentially the same after adjustment for parity, and when follow-up was censored at hysterectomy. The RR were 1.48 (95% CI=0.53–4.17) and 0.25 (95% CI=0.05–1.18), respectively, for the associations with uterine and cervical cancer mortality, and both RR were materially unchanged after adjustment for smoking and when follow-up was censored at hysterectomy. There was no association with vulvar/vaginal cancer .The RR suggested elevated risk of death from lung, urinary, and head and neck cancer, as well as non-Hodgkin's lymphoma, melanoma, and leukaemia, but confidence intervals were wide and included the null value. Further adjustment for smoking did not materially change the RR of the cancers known to be associated with that exposure. The inverse associations suggested for brain cancer, cancer of the upper GI tract, and ‘other’ cancer were also imprecise. The inverse association with death from multiple myeloma was statistically significant, RR=0.18 (95% CI=0.04–0.82), and was unchanged after adjustment for smoking. The data provided no evidence that DES was related to death from colorectal or other GI cancers.Our results are based on the largest mortality study to date of women with documented exposure to DES during pregnancy, with over 2000 deaths, of which 800 were from cancer and nearly 200 from breast cancer.Our data suggest that DES is associated with a small elevation of all-cause mortality. The finding may reflect residual confounding by a lifestyle correlate, although the RR was somewhat greater in the Dieckmann cohort, which consists of women who were enrolled in a clinical trial of DES and who received high doses of DES. There was no increase of cerebrocardiovascular deaths, and the suggested increased risk of vascular deaths was consistent with chance. Clinical trial results from the Women's Health Initiative (WHI) suggest increased risks of cerebrocardiovascular events, including stroke and thromboembolic events, in postmenopausal women using equine oestrogens, but the effects may be limited to current or recent use . Also, WThe increased risk of breast cancer death observed in the combined cohort is consistent with the large follow-up study of DES in relation breast cancer mortality , and ourWomen participating in the Dieckmann trial received especially high doses of DES resulting in a cumulative dose of 11–12 g over the course of the pregnancy. Based on studies involving women treated during obstetrics care , and cause of death was ascertained for 97.6% of the exposed and 98.5% of unexposed decedents. Although incomplete ascertainment may have attenuated some associations, it seems unlikely that ascertainment would be associated with DES exposure; so our results should not be biased away from the null.In summary, our findings indicate that women given DES during pregnancy experienced a slight but statistically significant elevation in all-cause mortality after adjusting for covariates and omitting breast cancer deaths. Our data did not show an elevated risk of mortality from cancer overall or from gynaecological cancers. However, an increased risk of breast cancer mortality was evident, particularly in the Dieckmann women, who received high doses of DES during pregnancy.


Recent research suggests that intrauterine exposures, perhaps factors that influence birth weight and other indicators of fetal growth, may affect future breast cancer risk. Because birth weight shows seasonal variation in Sweden, we assessed whether risk for breast cancer is associated with month of birth. The analyses included all 115,670 women, born between 1858 and 1968, who were reported to the Swedish Cancer Registry in 1958-89 as having breast cancer. Poisson regression models were used to examine the data. After adjustment for seasonality of number of live births in the population at risk, a significant seasonal pattern was identified for women born between 1880 and 1920. Women born in June had a 5% higher risk of breast cancer than those born in December. By contrast, there was no evidence of birth seasonality among 440,948 women with cancer at other sites. Exposures relevant to breast cancer risk later in life are unlikely to be related to month of birth. Thus, prenatal or early post-natal factors influence breast carcinogenesis, but the seasonal variation in these factors must have decreased over time.


The interval between successive births  may affect breast cancer risk, whereas interval from last birth to cancer onset may modify its behaviour.The study cohort consisted of 29 488 Finnish grand multiparous (GM) women, including 628 women with breast cancer. Conditional logistic regression for case–control design nested within the cohort was used to estimate proportional hazards . Age at first birth and parity were co-variables.Short interval (<1 year) between first and second birth increased the risk of advanced ductal breast cancer at ages < 50 years  as compared to interval 3+ years. The risk of advanced ductal cancer was also large  shortly (<3 years) after last birth as compared with the period 15+ years.Short birth interval-associated excess breast cancer risk may be related to stimulatory effects of female steroid hormones produced during two closely connected pregnancies, or defective breast maturation owing to failures in breastfeeding. Findings in our previous study of grand multiparous women  indicated that birth interval (time between two successive births) might affect the risk of breast cancer . The lenPregnancies have dual effects on breast cancer: an immediate increase in the risk after childbirth is followed by a long-term protection .This study population comprises of only GM women, most of whom belong to the religious minority, the Laestadian movement within the Lutheran church, which forbids the use of artificial contraception. Hence conception in this population takes place in physiological circumstances, thereby representing a specific advantage for our study. Here, we aimed to explore the impact of individual birth intervals on breast cancer risk at different stages. Another aim was to assess how the relative risk changes in relation to time since last birth in a Finnish cohort of GM women.The data of the Finnish national Population Register comprised 29 488 women born 1935 or later and registered as having at least five biological children by the end of 1997. This database involves complete linkages between parents and the children born in October 1953 or later, but precluding our obtaining full parity history for women born before 1935. The exactly registered birthdays of each child form the basis for birth interval calculations.Breast cancer cases diagnosed among the GM women between the fifth childbirth and 31 December 2006 were identified in an automatic record linkage with the files of the national population-based Finnish Cancer Registry, using personal identifiers as the key. There were 628 such breast cancers. The Finnish Cancer Registry files include clinical stage and cancer morphology. The registry receives cancer notifications from hospitals, pathological and cytological laboratories, and also from physicians outside hospitals; its coverage is almost 100% . Possible interactions were evaluated using the TPHREG procedure, a test release of the PHREG procedure that incorporates the CLASS statement in SAS 9.1. No interactions were found between the study variables.The RRs were also calculated for ductal and lobular, and for local and advanced  breast cancers separately. Further stratification was based on the age at the diagnosis of cancer (<50 years and 50+ years). Pregnancy at the age of 50 years or later is extremely rare. The younger women represent thus the years of reproduction and the older ones that of ovarian quiescence. The distribution of patients into different subcategories is presented in The likelihood ratio test was used to evaluate statistical significance of the parameters of interest. The trend test was analysed using linear trend test for the classified variables. The lowest category had value 1 and the highest category had the same value as the number of classes of the respective variable. Each of the four intervals between first and fifth births was classified into three categories . Parity  and the age at first birth  were added in each model. The time from the latest birth to the date of onset of cancer was stratified into four categories .Overall, the RR did not differ significantly between different birth intervals, whereas short interval (<3 years) between last birth and cancer onset was associated with a significantly increased risk as compared to those with the interval of 15+ years . IncreasFor cancers diagnosed before 50 years of age, the RR for short interval (<1 year) between first and second birth was increased twofold when compared with birth interval of 3+ years, and the interval from last birth to cancer onset among the cases was three times more often short (<3 years) than among the controls . An incrThe most significant findings appeared in advanced ductal cancer diagnosed before age 50 years . Short .This study finds that among GM women, breast cancer is associated with birth interval. Most important was the new finding that short birth interval between first and second birth was significantly associated with increased risk of advanced ductal cancer in young GM mothers. The risk of ductal breast cancer in clinically advanced stage is also high during the 3 years following the last birth.Previous studies on the risk factors in breast cancer have comprised predominantly nulliparous women and women with few children. We widened the perspective by evaluating this topic in a homogenous group of Finnish GM women, whose breast cancer risk is low, 45% having below the average incidence  as compared to 3+ years. The mechanism of this detrimental effect is unclear. The fact that it appeared only after the first pregnancy might reflect an association with breast maturation, because first pregnancy may have a central role in the differentiation of breast cells more resistant to carcinogenic influences (Theories of hormonal mechanisms of breast cancer development are primarily based on exposure to excessive endogenous oestrogens . The infStudies on the possible role of prolactin in breast cancer have yielded inconsistent results – protective or carcinogenic effects or no effect (In spite of our lack of breastfeeding data, we hypothesise that GM women, with increased breast cancer risk in association with a short interval between first and second birth, do not have breastfeed or have only for a very short time. Short-term breastfeeding, and especially no breastfeeding is associated with shorter birth interval . An incrBecause lactation participates in the differentiation of mammary epithelium in its terminal phase , deficieThe picture of breast cancer risk increase associated with birth intervals appeared to be more complicated than the previous one that long birth interval solely would affect the risk . In the The RR of breast cancer of young GM women was highest shortly after their last birth, which is in accordance with earlier observations of mothers with few pregnancies . TransieYoung age at first birth and increasing parity are established protective factors in breast cancer risk (Elevated risk of advanced ductal breast cancer of young GM women with short interval between first and second birth might have an association with female steroid hormones and prolactin secreted during two closely consecutive pregnancies or with defective maturation of the breast owing to inappropriate nursing.


A significant decrease in premenopausal breast cancer risk was also observed for increasing consumption of isoflavones . The present study found a statistically inverse association between tofu or isoflavone intake and risk of breast cancer in Japanese premenopausal women, while no statistically significant association was evident with the risk among postmenopausal women.Components of the Japanese diet, which might contribute to the relatively low breast cancer incidence rates in Japan, have not been clarified in detail. Since soybean products are widely consumed in Japan, a case–control study taking account of the menopausal status was conducted using data from the hospital-based epidemiologic research program at Aichi Cancer Center (HERPACC). In total, 167 breast cancer cases were included and 854 women confirmed as free of cancer were recruited as the control group. Odds ratios (OR) and 95% confidence intervals (95% CI) were determined by multiple logistic regression analysis. There were reductions in risk of breast cancer associated with high intake of soybean products among premenopausal women. Compared with women in the lowest tertile, the adjusted ORs for top tertile intake of  Despite a marked increase in recent years, the incidence rates for female breast cancer in most Asian countries are much lower than those in the Western world . Recentltofu (soybean curd), okara (tofu lees), moyashi (soybean sprouts), tonyu (soymilk), yuba (soy milk skin), kinako (soy flour), miso (fermented soybean paste), atsuage (deep fried tofu), aburage (thinly sliced deep fried tofu), natto (fermented soybeans), koyadofu (freeze dried tofu) and shoyu (soy sauce), so that the diet is likely to be much richer in isoflavones than in the Western world, where the diet usually does not include soybeans. We here evaluated the association between risk of breast cancer and consumption of soybean products and isoflavones using data from the hospital-based epidemiologic research program at Aichi Cancer Center (HERPACC).Soybeans provide a unique concentrated source of isoflavones and soybeans or isoflavones have been shown to exert anticarcinogenic effects on hormone-related cancers in a large number of experimental studies. Despite the growing interest in the protective effects, there are relatively few epidemiological data available since soybean products are consumed mainly by Asian populations. In Japan, intake is in various forms, including Details of the study design and subject characteristics have been described elsewhere , 2001. IThe questionnaire included questions on occupation, medical history, height, weight, weight at around 20 years of age, family history (parents and siblings), smoking and drinking habits, sleeping habits, physical exercise and reproductive history. The details were taken prior to assessment of symptoms and all information was collected before clinical diagnoses were made.The present analysis was restricted to women aged 30 and over who visited hospitals between January 2001 and March 2002. As eligible cases, 79 premenopausal and 88 postmenopausal women diagnosed as having breast cancer by histological examination within 6 months of the first-visit were recruited. A methodological study applying the same HERPACC data set earlier showed that the OR based on a large number of controls gives more power and steadier estimates than the use of matched controls  (tofu (soybean curd); (b) miso (fermented soybean paste); (c) atsuage (deep fried tofu); (d) aburage (thinly sliced deep fried tofu); (e) natto (fermented soybeans); (f) koyadofu (freeze dried tofu). We evaluated validity of intakes based on the questionnaire against those according to 28-day (four-season consecutive 7-day) weighted diet records among 79 Japanese female dietitians. In the validation study, the Spearman’s correlation coefficients between the estimate intake of soybean products and isoflavone from the questionnaire and that from dietary records were 0.51, 0.53, respectively. For reproducibility of estimation from the questionnaire, the Spearman’s correlation coefficients for the consumption of soybean products and isoflavone intake between two questionnaires administered 1 year apart were 0.57 and 0.47, respectively.Furthermore, the following soybean products were included: (a) http://www.nal.usda.gov/fnic/foodcopm/Data/isoflav/isoflav.html). Isoflavone intake was calculated using consumption of six items of soybean products, green soybean, peanuts, Japanese green tea, and vegetables other than green-yellow vegetables such as cucumber, eggplant, lettuce, bean sprouts, onion, Chinese cabbage.All subjects in the present study were asked for average frequency and portion size of consumption, during the period of 1 year before onset of the present disease or before the interview. There were eight categories of possible responses, ranging from‘rarely or never’ to ‘three or more times per day’. For each food, a commonly used unit or portion size was specified and the interviewers asked the subjects using sampling models of full-scale photographs. We ascertained average amount of daily consumption of each food and nutrients by multiplying the food intake (in grams) or serving size and the nutrient content per 100 grams of food as listed in the Standard Tables of Food Composition, Version 5 and the Follow-up of Standard Tables of Food Composition , 2001. It-test and all P-values presented are two-sided.Dietary intake data were analysed by individual food items, food groups and nutrients for all subjects combined and separately for premenopausal and postmenopausal women. The differences of means were examined by P-value for trend corresponded to the estimate of the slope derived from the logistic model in the case that the integers, 1 to n, were assigned to the ordered n levels of each factors. The LOGISTIC procedure provided by SAS  was utilised to perform the calculations. To compare differences and similarities of effects of Japanese diet on risk for breast cancer, subjects were stratified with reference to menopausal status. The ORs of breast cancer are calculated for the selected food groups and isoflavone intake in tertiles, with the lowest tertile as the referents. Multivariate models was adjusted for age, energy (as a continuous variable), motives for consultation , smoking status , drinking status , exercise , family history (yes or no), age at menarche , parity , age at first full-term pregnancy , and current body mass index (BMI) . Further, inclusion in the model of age at menopause  was performed with the calculations for postmenopausal women. Energy was adjusted for the multivariate nutrient density method.Logistic regression analysis was used to obtain odds ratios (ORs) and 95% confidence intervals (95% CI) as estimates of relative risk. The 2 (kg m−2), according to Quetelet's formula and current BMI values were stratified into three categories. Since a BMI⩾25 is defined as obese by the Japanese Society for the Study of Obesity, the cutoff for the highest BMI group was BMI⩾25.A positive family history may involve different types and numbers of relatives. Distant relatives share less genetic influence and fewer confounding environmental and/or behavioral factors than do close ones. Furthermore, information on the medical history of distant relatives is limited and less precise than that of close or first-degree relatives. In this study, the presence of either a mother or sister with breast cancer was considered as a positive family history. BMI was calculated as weight/heightP<0.04). Among premenopausal women, the means of soybean products were 51.7 g  in case group and 63.5 g  in control group . Breast cancer cases reported lower total isoflavones intake per day than controls, with averages of 20.8 mg  and 25.8 mg , respectively  among premenopausal women. We observed no other significant differences among premenopausal women and there were no significant differences between cases and controls in average intake of food groups and nutrients among postmenopausal women.P=0.06). Among postmenopausal women, on the other hand, consumptions of soybean products, meat and meat products, fish and fish products, vegetables, fruits, dairy products were not associated with the risk of breast cancer.Among premenopausal women, breast cancer risk was inversely associated with consumption of soybean products . The ORstofu consumption and breast cancer in premenopausal women. ORs were 0.44 , 0.49  for the second to the top tertiles of intake compared with the lowest tertile of intake of tofu . Compared with those in the lowest tertile of atsuage (deep fried tofu) consumption, the adjusted OR for breast cancer in top tertile was 0.70  among premenopausal women. On the other hand, a significantly increased risk of breast cancer with consumption of atsuage was observed in postmenopausal women. The adjusted OR for top tertile of atsuage consumption was 2.28 . Also, consumption of aburage (thinly sliced deep fried tofu) showed a similar positive association with the risk of breast cancer. There were no associations with intake of tofu, miso and natto intake among postmenopausal women. With intake of koyadofu (freeze dried tofu) divided into tertiles, there was no apparent modification of the breast cancer risk among either pre- or postmenopausal women.The ORs for breast cancer according to type of soybean product are presented in P=0.02). Among postmenopausal women the similar trend was also observed; however, there was no statistically significant association between consumption of isoflavone and breast cancer risk  or isoflavones intake in Japanese premenopausal women while there was no statistically significant link among their postmenopausal counterparts.The present study found a statistically inverse association between risk of breast cancer and soybean products , making it impossible to investigate differences in effects among types of soybean-ingredient foods.Soybean foods are rich in precursors of the isoflavone daidzein and genistein, which are heterocyclic phenols similar in structure to oestrogens, and it has been hypothesised that a high dietary intake of soybean products might reduce breast cancer risk by interfering with the action of endogenous oestradiol . The resTofu, miso and natto were main foods containing rich isoflavones. Attributable rate of genistein were tofu (49.6%), miso (20.9%), natto (14.7%) among Japanese. In the present study, tofu was protective in premenopausal women, while atsuage and aburage, deep fried tofu containing much oil, was associated with elevated risk among postmenopausal women. This may be due to fat intake, which can exert an influence on the development of breast cancer among postmenopausal women. Some studies have suggested that high intake of soybean products in premenopausal women may reduce serum oestradiol concentrations, suppress the mid-cycle surge of gonadotropins, and perhaps increase the length of the menstrual cycle  increase of serum levels of sex hormone-binding globulin (SHBG) ; (2) dowe type I . StudiesSoybeans are a unique dietary source of a group of phytochemicals and several natural anticarcinogens have now been identified in soybeans, such as protease inhibitors, phytates, phytosterols, saponins and lignans. Also, soybeans are an excellent source of dietary fibre and micronutrients, especially calcium. It will be important in future epidemiological studies to investigate the association between intake of soybean products and breast cancer by obtaining a more complete assessment of soybean intake. Further studies are required to confirm the ability of the nutritional profile of soybeans to reduce the risk of breast cancer.Potential limitations of the present study should be considered. One methodological issue is selection of base population for controls. We applied noncancer patients at ACCH as controls because it is reasonable to assume our cases arise within this population base. Main reasons to visit at ACCH among cases and controls were self/family recommendation, referral from other clinics, and secondary screening after primary screening. Although distributions of these reasons slightly differ, it was considered in statistical analyses. Notable point of our control population is its similarity to general population in terms of exposure of interest, here dietary pattern. We have compared lifestyle characteristics between outpatients in ACCH and the 1231 individuals randomly selected from the general population, and confirmed that they are not substantially different . Much larger studies will be required to confirm the impact of soybean products and isoflavone intake on breast cancer risk among Japanese women.Another limitation of the present study included the small number of cases that are more likely to be due to chance leading to a false-positive result. Conversely, small sample may lack sufficient power to detect significant differences. The powers for detection of the OR of 0.5 for higher A number of risk factors for breast cancer have been established, most of them related to reproductive events. Evidence from studies of migrant populations, however, has also implicated environmental or lifestyle factors as being of importance and the diet is suspected of playing a role. The present study focused on the Japanese diet, which features a high level of consumption of soybean products and was able to demonstrate that high consumption of soybean products reduces the risk in Japanese premenopausal women. The findings are biologically plausible and suggested a potential beneficial effect of soybean products and isofloavones in the prevention of breast cancer.


CASP8 at 2q33.1 have been associated with the risk of developing cancer, specifically, the D302H variant (rs1045485) with breast cancer in the European population and the −652 6N ins/del promoter variant (rs3834129) with multiple tumours including colorectal cancer (CRC) in the Chinese population. We evaluated the relationship between −652 6N ins/del and D302H variants and risk of developing CRC in the UK population by genotyping 4016 cases and 3749 controls. Both variants showed no evidence of an association with risk of developing CRC . In contrast, the recently identified CRC susceptibility allele rs6983267 mapping to 8q24 was significantly associated with disease risk (P=8.94 × 10−8). It is thus very unlikely that variation in CASP8 defined by −652 6N ins/del or D302H influences the risk of CRC in European populations. The implications of our findings both in terms of population-specific effects and publication bias are discussed.Polymorphisms in  Although inherited susceptibility is responsible for ∼30% of all colorectal cancers (CRCs)  maps, and consequently of sets of tagging SNPs that capture much of the common sequence variation, allows whole-genome-wide association studies (GWAS) for disease associations to be efficiently conducted. This approach is attractive as it is unbiased and does not depend upon prior knowledge of function or presumptive involvement of any gene in disease causation. Furthermore, it minimises the possibility of failing to identify important variants in hitherto unstudied genes. Recent GWAS have vindicated the assertion of the ‘common-disease common-variant’ hypothesis providing evidence for common disease variants for CRC mapping to 8q24  is a key regulator of apoptosis or programmed cell death, an essential defense mechanism against hyperproliferation and malignancy. Polymorphic variation in CASP8 has been reported to influence cancer risk. The variant D302H (rs1045485) has been associated with risk of breast cancer in the European population , the Genetic Lung Cancer Predisposition Study  and the Royal Marsden Hospital Trust/Institute of Cancer Research Family History and DNA Registry   were ascertained through the National Study of Colorectal Cancer Genetics (NSCCG) . Genotyping of D302H (rs1045485) and −652 6N ins/del (rs3834129) was conducted by competitive allele-specific PCR KASPar chemistry . For rs1045485, allele-specific primers were: GAAGGTGACCAAGTTCATGCTAGATTTGCTCTACTGTGCAGTCATC and GAAGGTCGGAGTCAACGGATTAGATTTGCTCTACTGTGCAGTCATG; common primer: GACCACGACCTTTGAAGAGCTTCAT. For rs3834129, allele-specific primers were: GAAGGTCGGAGTCAACGGATTGCCATAGYAATTCTTGCTCTGCCAA and GAAGGTGACCAAGTTCATGCTGCCATAGYAATTCTTGCTCTGCCAC; common primer: CACTGAGACGTTAAGTAACTTGCCCAA. Genotyping quality control was tested using duplicate DNA samples within studies and SNP assays, together with direct sequencing of subsets of samples to confirm genotyping accuracy. For both SNPs, 99.6% concordant results were obtained.P-value of 0.05 was considered significant. Linkage disequilibrium between genotypes was enumerated using the metric R2. This was estimated using Haploview software.Statistical analyses were undertaken using STATA Software . To test for population stratification, the distribution of genotypes in controls was tested for a departure from Hardy–Weinberg equilibrium (HWE). Logistic regression was used to calculate odds ratios (ORs) and the associated 95% confidence intervals (CIs). Power calculations were undertaken on the basis of proportions. In all analyses, http://www.ncbi.nlm.nih.gov/projects/SNP/)). In the Chinese population, this SNP has, however, been documented to be nomomorphic. The frequency of −652 6N del allele in controls was 0.501, again similar to previously published data on Europeans. It is, however, markedly different to that observed in the Chinese population in which the frequency is ∼25%.D302H (rs1045485) and −652 6N ins/del (rs3834129) genotypes were obtained for 96% (3843) and 97% (3879) of CRC cases and 97% (3631) and 98% (3661) of controls, respectively; hence, there was no evidence of any systematic bias in genotyping. Furthermore, there was no evidence of population stratification, as the genotype distribution in controls for both variants satisfied HWE . The freP=0.423 and 0.216 respectively; r2=0.19). By comparison, the 8q24 variant rs6983267 displayed a strong association with CRC risk (P=8.94 × 10−8).The frequencies of both variants were not significantly different in cases and controls (allele test of significance, P-value of 0.05. Hence, while the D302H polymorphism has been robustly shown to influence breast cancer risk (Based on the number of cases and controls analyzed and the population frequencies of the rare alleles of each polymorphism, our study has ∼90% power to demonstrate an OR of 1.2 associated with each variant stipulating a The −652 6N ins/del variant has been reported to influence both breast and CRC in the Chinese population. A recent study of breast cancer with the Europeans has failed to confirm an association with breast cancer risk (et al, used PCR-RFLP to determine genotypes, which through incomplete digestion of PCR product can lead to misassignment of alleles. This can lead to systematic basis, especially if cases and controls are not genotyped simultaneously. Although we cannot entirely exclude the possibility of such random measurement error as the basis of non-replication of study findings, it seems unlikely to be a major issue.In our study, genotyping was conducted by allele-specific PCR, generally acknowledged to provide a robust means of assigning genotypes. In contrast, Sun While there is evidence that difference in genetic frequencies across populations of different racial origin is not generally accompanied by differences in population-specific genetic effects (in vitro functional assays, it was asserted that this variant has direct impact of cancer risk via apoptosis. Allele frequencies are, however, very different between populations and the possibility of differential LD in the different populations remains a basis for non-replication. Alternatively, as the study purporting that −652 6N ins/del influences CRC was based on a relatively small sample size to that advocated for GWA analyses, there remains the distinct possibility of a type 1 error coupled with publication bias. If additional studies of these variants are conducted, it opens up the possibility of using meta-analysis to provide the final step in confirming or refuting an association with cancer risk.On the basis of


Gender of offspring is influenced by maternal hormonal level during pregnancy, which is blieved to influence the subsequent maternal breast cancer risk. However, analysing national birth and cancer registrations in a cohort of 998 499 women, we found no association between gender of offspring and subsequent breast cancer risk. © 2000 Cancer Research Campaign


P<0.00001) for an intake of 35–44 g per day alcohol, and 1.46  for ⩾45 g per day alcohol. The relative risk of breast cancer increased by 7.1%  for each additional 10 g per day intake of alcohol, i.e. for each extra unit or drink of alcohol consumed on a daily basis. This increase was the same in ever-smokers and never-smokers . By contrast, the relationship between smoking and breast cancer was substantially confounded by the effect of alcohol. When analyses were restricted to 22 255 women with breast cancer and 40 832 controls who reported drinking no alcohol, smoking was not associated with breast cancer . The results for alcohol and for tobacco did not vary substantially across studies, study designs, or according to 15 personal characteristics of the women; nor were the findings materially confounded by any of these factors. If the observed relationship for alcohol is causal, these results suggest that about 4% of the breast cancers in developed countries are attributable to alcohol. In developing countries, where alcohol consumption among controls averaged only 0.4 g per day, alcohol would have a negligible effect on the incidence of breast cancer. In conclusion, smoking has little or no independent effect on the risk of developing breast cancer; the effect of alcohol on breast cancer needs to be interpreted in the context of its beneficial effects, in moderation, on cardiovascular disease and its harmful effects on cirrhosis and cancers of the mouth, larynx, oesophagus and liver.Alcohol and tobacco consumption are closely correlated and published results on their association with breast cancer have not always allowed adequately for confounding between these exposures. Over 80% of the relevant information worldwide on alcohol and tobacco consumption and breast cancer were collated, checked and analysed centrally. Analyses included 58 515 women with invasive breast cancer and 95 067 controls from 53 studies. Relative risks of breast cancer were estimated, after stratifying by study, age, parity and, where appropriate, women's age when their first child was born and consumption of alcohol and tobacco. The average consumption of alcohol reported by controls from developed countries was 6.0 g per day, i.e. about half a unit/drink of alcohol per day, and was greater in ever-smokers than never-smokers, . Compared with women who reported drinking no alcohol, the relative risk of breast cancer was 1.32  87, 1234–1245. doi:10.1038/sj.bjc.6600596www.bjcancer.com© 2002 Cancer Research UK The published results from these studies have generally suggested that women who regularly consume alcohol may be at a slightly increased risk of the disease, but the findings reported for tobacco are inconsistent. Alcohol and tobacco consumption are known to be associated one with another, and published results have not always allowed adequately for possible confounding between these exposures. Individual data from 65 epidemiological studies of breast cancer 63 published6263 and two unpublished in which information on alcohol and/or tobacco consumption had been collected contributed to this collaboration. These studies, some of which have not published results for alcohol or tobacco, include over 80% of the worldwide information on the topic (see Appendix II (www.bjcancer.com)). The data from these studies were analysed, taking careful account of the possible confounding between alcohol and tobacco consumption, as well as confounding by other factors.Many epidemiological studies have investigated the relationship between breast cancer and the consumption of alcohol and/or tobacco. References to over 80 studies that have collected relevant data, as well as to reviews of the subject, are given in Appendix II . No information was sought about alcohol consumption at various ages or about the particular type of alcohol consumed. Information was also sought on whether or not each woman had ever smoked, and whether she was a current or past smoker. Active smoking only was considered and no attention given to the reported associations with environmental tobacco smoke,49 nor was information sought on the age women were when smoking started or stopped, or on the amount smoked. The methods of identifying studies and of data checking, and correction, have been described elsewhere.65Some investigators provided estimates of alcohol intake reported by each woman expressed as gram (g) of alcohol consumed per day or per week. Others provided information on the reported number of alcoholic drinks consumed daily or weekly. In such instances, the number of grams of alcohol consumed per day, was estimated assuming that one alcoholic drink contains 12 g alcohol in the USA and Italy,6667 Data from different studies were combined by means of the Mantel–Haenszel stratification technique, the stratum-specific quantities calculated being the standard ‘observed minus expected’ (O–E) numbers of women with breast cancer, together with their variances and covariances. These values yield both statistical descriptions  and statistical tests . When only two groups are being compared, relative risk estimates are obtained from O–E values by the one-step method,66 as are their standard errors (SE) and confidence intervals (CI). When more than two groups are compared, variances are estimated by treating the relative risks as floating absolute risks (FARs).67 This approach yields floated standard errors (FSE) and floated confidence intervals (FCI). Presentation of the results in this way enables valid comparisons between any two exposure groups, even if neither is the baseline group. Any comparison between groups must take the variation in each estimate into account by summing the variances of the logarithms of the two FARs.Statistical methods were similar to those used in previous reports by this group.To obtain comparability between the women with breast cancer and similar women without breast cancer, all analyses were routinely stratified by study, and centre within study; by age ; by parity and, where appropriate, age when the first child was born  and age at first birth ). Where appropriate analyses relating to alcohol consumption were stratified by smoking history (ever/never) and analyses relating to tobacco consumption were stratified by alcohol consumption . In order to summarise the relationship between alcohol consumption and breast cancer risk, a linear trend in the log relative risk of breast cancer was fitted across increasing categories of consumption. In estimating such trends, the median consumption within a given category was taken to be the level of alcohol consumption for that category.In general, results in the text are presented as relative risks and their appropriate SE or FSEs. Where results are presented in the form of plots, relative risks and their corresponding CIs/FCIs are represented by squares and lines, respectively. The position of the square indicates the value of the relative risk and its area is inversely proportional to the variance of the logarithm of the relative risk, thereby providing an indication of the amount of statistical information available for that particular estimate. Owing to the large number of relative risk estimates calculated, results are given with their appropriate 99% CIs/FCIs; and 95% CIs/FCIs are used only to summarise the main findings.65 assuming that an intake of 10 g per day is roughly equivalent to one unit or drink of alcohol per day. The cumulative incidence of breast cancer up to age 80 years was calculated from the age-specific findings.The absolute risk of breast cancer associated with various levels of alcohol consumption was estimated for women in developed countries, by applying the dose-response estimates obtained here to age-specific incidence rates for breast cancer in developed countries around 19906263 and two unpublished studies. Information on both alcohol and tobacco had been collected in 53 of these studies, that included a total of 58 515 cases and 95 067 controls from 51 published51 and two unpublished studies. Unless otherwise specified, analyses presented here are restricted to data from these 53 studies. This enables women to be cross-classified by both their alcohol and tobacco consumption, thus permitting adequate examination of possible confounding between the two exposures.The 65 studies that contributed individual data on alcohol and/or tobacco consumption and other factors relevant to breast cancer included a total of 66 426 women with invasive breast cancer (cases) and 126 953 women without breast cancer controls from 63 published48 were conducted in developed countries. Among controls, alcohol consumption was substantially greater in women from developed than developing countries . The proportion of controls from developed countries who reported drinking no alcohol was 40%, and a further 28% reported consuming <5 g per day, i.e. less than half a unit/drink of alcohol per day .Overall about half the women in developed countries reported that they had ever smoked, but smoking habits varied considerably according to alcohol intake, both for cases and controls . Among cBecause alcohol and tobacco consumption are so closely associated, analyses of their effects were initially carried out separately for never-smokers and ever-smokers  and for drinkers and non-drinkers (in the case of tobacco).P<0.00001 in each group). The trends associated with increasing levels of alcohol intake in never-smokers and ever-smokers did not differ significantly from each other . Therefore subsequent analyses concerning alcohol consumption include both never-smokers and ever-smokers, and the data are stratified by smoking history as well as by study, age, parity and age at first birth.Table 2P<0.00001) for each additional 10 g per day intake of alcohol, i.e. for each extra unit/drink of alcohol consumed on a daily basis  for women whose reported alcohol consumption was 35–44 g per day and 1.46  for a consumption of ⩾45 g per day, where the average consumption was 57 g per day.When the data in smokers and non-smokers were combined the relative risk of breast cancer increased with alcohol intake, increasing by 7.1%  or according to study design . In the one study52 which contributed data on alcohol, but not smoking, the estimated increase in the relative risk of breast cancer per additional 10 g per day intake was 13.8% (SE 10.5%). Because of the large standard error, the estimated increase in relative risk in this study does not differ significantly from results for all other studies combined .The study-specific results are summarised in Figure 2The effect of adjusting for 11 other potential confounding factors  on the relationship in Table 3vs never-smokers=1.03, SE 0.023; NS). However, among women who reported drinking alcohol, the findings for smoking were difficult to disentangle from the effects of the alcohol itself. When ever-smokers were compared to never-smokers the relative risk for breast cancer was 1.09 (0.018) before stratification by the amount of alcohol consumed, and declined to 1.05 (0.020) after stratification. The corresponding χ21 value declined by three-quarters from 23.4 to 6.4. Since alcohol consumption is known to be unreliably measured,68 and stratification for such a poorly measured variable reduced the χ2 value by three-quarters, stratification by true alcohol intake would be expected to reduce the χ2 value by even more.69 Since it is not possible to eliminate residual confounding among drinkers, results concerning tobacco consumption are restricted to women who reported drinking no alcohol at all, where such confounding should be minimised.Among the 22 255 cases and 40 832 controls who reported drinking no alcohol, the risk of breast cancer in ever-smokers did not differ significantly from that in never-smokers  or study design . Information on current and past smoking was available for all but five studies.28 (and two unpublished). Among ever-smokers in the remaining 48 studies 54% were current smokers and 46% were past smokers. Compared to never-smokers the relative risk of breast cancer was 0.99 (SE 0.03) for current smokers (Appendix III in www.bjcancer.com)), and 1.07 (SE 0.03) for past smokers (Appendix IV in www.bjcancer.com)).The study-specific relative risks for breast cancer in ever-smokers compared to never-smokers are shown in Figure 3vs 7.6%) or a hysterectomy without bilateral oophorectomy (13.3% vs 12.5%). The average age at bilateral oophorectomy was 41.6 (SD 7.5) and 44.2 (SD 6.6), respectively and the average age at hysterectomy was 38.6 (SD 9.3) and 40.0 (SD 9.9), respectively. Average age at natural menopause was also slightly earlier in ever-smokers than in never smokers, at 48.3 (SD 4.8) and 49.3 (SD 4.7) years, respectively. The relative risk of breast cancer in ever vs never-smokers was similar for women who had had an oophorectomy, hysterectomy or natural menopause , but because of the potential for confounding by alcohol the results from these studies have not been included in the main analyses.Eleven studies215=18.0; P=0.3). Nor was there significant variation in the relative risk of breast cancer associated with having ever smoked across categories of the 15 characteristics examined .The increase in the relative risk of breast cancer for each additional 10 g per day intake of alcohol consumption was calculated separately for various subgroups of women, subdivided according to 15 personal characteristics including their age, childbearing pattern, race and familial patterns of breast cancer. Overall there was no significant variation in the trend associated with increasing intake of alcohol between categories defined by any of the 15 factors examined . There was no significant difference in the extent of tumour spread among the cases according to tobacco consumption .Information on the extent of spread of the breast cancer was available for about 60% of the study population. Both for tumours localised to the breast and for tumours that had spread beyond the breast, the risk of breast cancer increased with increasing alcohol consumption (increase in relative risk of breast cancer of 6.9% (1.3%) and 9.4% (1.5%), respectively, per 10 g per day alcohol consumption: χ70 The average consumption of alcohol by controls studied here from developed countries was 6.0 g per day. If the dose-response relationship described here is valid, it is estimated that about 4% of breast cancers in developed countries are attributable to alcohol. The cumulative incidence of breast cancer by age 80 years is estimated to increase from 8.8 per 100 women in non-drinkers to 9.4, 10.1, 10.8, 11.6, 12.4 and 13.3, respectively, per 100 women consuming an average of 1, 2, 3, 4, 5 and 6 alcoholic drinks each day . The observed association between breast cancer and alcohol consumption is therefore unlikely to be an indirect effect of tobacco.The relative risk of breast cancer was found to increase with increasing intake of alcohol, both in never-smokers and in ever-smokers, and the magnitude of the increase was the same in each group ), but because results were presented in a different way in each study, it is difficult to combine the published data directly. Nevertheless, out of the six largest studies all but one ) reported a statistically significant increased risk of breast cancer with increasing intake of alcohol. Each of these six studies included at least 500 women with breast cancer and altogether they comprised most of the information that had not been contributed in this collaboration. The remaining 14 studies were comparatively small and none of their published results on alcohol differed substantially from those reported here. Therefore the findings on alcohol and breast cancer from studies not included here do not appear to differ materially from these results.As far as can be ascertained, over 80% of the worldwide epidemiological data that have been assembled on the relationship between breast cancer and alcohol and tobacco consumption were contributed to this collaboration. Another 20 studies were identified with relevant data that together included about 12 000 women with breast cancer (see Appendix II in www.bjcancer.com)). None of these studies has, however, published results on the risk of breast cancer in relation to smoking, restricted to women who never drank alcohol.Only one of the 20 studies that had not contributed to this collaboration claimed that smoking is associated with an increased risk breast cancer . However, self-reported information on alcohol consumption is known to underestimate true consumption.68 Systematic under-reporting of consumption by both cases and controls would result in an overestimation of the relative risk of breast cancer for a given level of alcohol consumption. By contrast, random misclassification among both cases and controls would have the opposite effect, resulting in an underestimation of the relative risk. These two types of measurement error are inevitable, but counter-acting, and it is not possible to estimate their overall effect on the relative risks calculated here. Moreover, the shape of the dose-response relationship could be changed if, for example, heavy drinkers were more likely to under-report intake than moderate drinkers. Taken together, these reporting errors imply that some uncertainty remains about the true quantitative effect of an intake of a fixed amount of alcohol on the risk of developing breast cancer.Overall, the relative risk of breast cancer appeared to increase by 7.1% (95% CI 5.5–8.7%) for each additional 10 g per day intake of alcohol i.e. for each extra unit/drink of alcohol consumed on a daily basis. Information on alcohol consumption was, however, usually self-reported, describing drinking habits at around the time that the women were interviewed. No information on the pattern of intake, including the type of alcohol consumed and the duration of intake, was collected for this collaboration. There is no strong evidence here to suggest biased reporting of alcohol consumption in case-control studies, since there was no significant difference in results between case–control and cohort studies ). The findings from case–control studies could, in theory, be biased if women with breast cancer stopped smoking when they first developed symptoms, or if there were differential reporting of smoking by cases and controls. However, the results from cohort studies, where exposure information was collected prospectively, suggest no increase in the risk of breast cancer in ever-smokers or current smokers compared to never-smokers .With respect to the consumption of tobacco, the main exposure variable examined here was whether or not a woman had ever smoked. No information was collected for this collaboration on the amount smoked or on the age that smoking started or stopped, nor has attention been given to the reported effects of environmental exposure to tobacco,74If the pattern of breast cancer associated with increasing levels of alcohol consumption estimated here is valid, then about 4% of the breast cancers in women in developed countries may be attributable to alcohol. The consumption of alcohol by most women in developed countries is relatively low, with about two-thirds consuming little or no alcohol each day. For women in developed countries who regularly drink alcohol, the lifetime risk of breast cancer is estimated to increase by about 0.7 per 100 women for each extra unit or drink of alcohol consumed on a daily basis. For example, the cumulative incidence of breast cancer by age 80 years is estimated to increase from 8.8 per 100 women who drink no alcohol to 10.1 or 100 who consume two alcoholic drinks daily and to 11.6 per 100 who consume four drinks daily. This excess risk should be considered in the context of the beneficial effects of alcohol, in moderation, on cardiovascular disease, and its harmful effects on cirrhosis and on cancers of the mouth, larynx, oesophagus and the liver.