The involvement of vitamin D receptor (VDR), which is a key mediator in the vitamin D pathway, in breast cancer etiology has long been of interest.VDR gene, specifically BsmI and Poly(A), and breast cancer risk within a large, population-based, case-control study of breast cancer. Cases  were Caucasian and African-American women, aged 35 to 64 years, who were diagnosed with incident, invasive breast cancer between July 1994 and April 1998. Control individuals  were women without breast cancer ascertained through random digit dialing.We examined the association between polymorphisms in the 3' end of the BsmI . However, no associations with the bb genotype were observed in African-American women. Overall, there were no significant associations between the Poly(A) genotype and breast cancer risk in either racial group. Smoking status (ever/never) modified the association between both the BsmI and Poly(A) genotypes and breast cancer risk. The respective associations between these genotypes and breast cancer risk did not significantly vary by oral contraceptive use, hormone replacement therapy, or body mass index.Accounting for age, study site, and sampling weights, we observed a significantly increased risk for breast cancer among Caucasian, postmenopausal carriers of the bb genotype of BsmI bb genotype and shed light on possible differential effects by menopausal status and race.Our results provide additional support for an increased risk for breast cancer in postmenopausal Caucasian women with the  Vitamin D is a key player in cell proliferation, differentiation, and apoptosis in normal and malignant breast cells ,2. IndeeVitamin D has also been established as a determinant of bone density and, furthermore, higher bone density has been shown to be associated with risk for developing breast cancer -9. WhethThe autocrine/paracrine pathway of vitamin D biosynthesis has been implicated in breast cell carcinogenesis . In thisVDR gene exhibit sequence level variation in the population. Three restriction fragment length polymorphisms at the 3' end of the gene  have been observed to be in linkage disequilibrium with one another and with the variable length Poly(A) sequence in the 3'-untranslated region [VDR gene may modify the risk for breast cancer, either singularly or through gene-gene or gene-environment interactions.Both the 5' and 3' termini of the d region -15. BecaPoly(A) has been associated with both increased and decreased risks for breast cancer [Poly(A) tract has been associated with prostate cancer risk in a subset of studies [BsmI polymorphism and breast cancer risk have reported positive [BsmI polymorphism and bone density [The long sequence of t cancer ,16,17. A studies -14. Studpositive ,10,16, ipositive -20. More density ,22.VDR polymorphisms and breast cancer risk in African-American women. The present study was undertaken to assess whether VDR polymorphisms BsmI and Poly(A) are associated with breast cancer risk in a large population-based case-control study of Caucasian and African-American women aged 35 to 64 years.The inconsistency in findings to date is not surprising because most studies have been limited by sample size, and there has been considerable heterogeneity in study designs. Moreover, prior studies have focused on primarily Caucasian populations; hence, little is known about the association of This study was conducted within the National Institute of Child Health and Human Development's Women's Contraceptive and Reproductive Experiences (CARE) study , the detP < 0.001; controls: 68.6% versus 44.1%, P < 0.001), to have attended college , and to have local stage disease . Additionally, among only the controls, those who gave blood were more likely than those who did not to have a positive family history of breast cancer . The weighted sampling probabilities used in these analyses should alleviate these differences to some extent.Of interviewed women, 33% were sampled for blood collection based on available funding. The stratified sampling plan for blood collection consisted of a blood sample from all cases and controls with a first-degree family history of breast cancer  plus a random sample of those without a first-degree family history. Random selection for blood collection was based on sampling fractions specific to study center, race, and 5-year age group to achieve a relatively uniform distribution across strata. Of the 2,049 cases and 1,949 controls selected for blood draw, 1,644 (80.2%) cases and 1,451 (74.3%) controls donated a sample. This represents 35.9% and 31.0% of all CARE cases and controls, respectively. Among both cases and controls selected for blood collection, the proportions who gave blood did not vary by age. However, those who gave blood were proportionately more likely than those who did not give blood to be Caucasian  provided genotyping data on one or both of the VDR polymorphisms; the other 29 samples were excluded because they did not provide data on either polymorphism. Study participants provided written informed consent for the interview and for the use of specimens for laboratory analysis.BsmI VDR polymorphism is located within intron 8 of the gene. Initial PCR amplification involved 25 to 50 ng genomic DNA, purified from whole blood using standard proteinase K, phenol chloroform extraction methods [BsmI VDR polymorphism was determined using PCR-restriction fragment length polymorphism analysis, as we described previously [The  methods . The BsmBsmI at 65°C for 18 hours using 5 units of enzyme  per 20 μl reaction. Following digestion, the PCR products were separated using 2% agarose gels containing ethidium bromide and visualized under short-wave UV light. Fragments of 650 and 150 bp are visible if the 800 bp product is cut by the BsmI restriction enzyme.The resulting 800 base pair (bp) PCR product is then diluted and digested with BsmI restriction site appeared on the gel as single 800 bp band. DNA from homozygote individuals (bb) appeared as two well-separated bands, 650 and 150 bp, indicating that the BsmI enzyme site is present in both alleles. Heterozygotes (Bb) have three bands: a 650 bp band and a 150 bp band  plus an 800 bp band (indicating its absence in the other). Because the assay is extremely robust and it is easy to detect heterozygotes as well as both types of homozygotes, no additional quality control procedures were implemented except water blanks to check for sample contamination, standards to size restriction digestion products, and orientation markers.DNA from homozygote individuals (BB) lacking a VDR Poly(A) polymorphism was analyzed using a nested PCR reaction [The reaction . In the The PCR products were resolved on a Li-Cor Model 4200 automated infrared DNA sequencer . Genotypes were assigned with commercially available SAGA software (Li-Cor). Allele sizes are scored by comparison with known control sizes and confirmed by rerunning with additional samples of known size.Sampling weights were computed by dividing the total number of women interviewed in each of the 240 strata  by the number of women sampled for the genotyping. These weights allowed results from tested samples to be adjusted so that they represented the proportions and effects expected if the entire CARE study population had been tested, and were used to calculate weighted proportions and weighted odds ratios (ORs). These methods were described previously .2 test was used to check Hardy-Weinberg equilibrium and to compare case and control distributions across covariate categories. The correlation coefficient r was used as the estimate of pair-wise linkage disequilibrium between BsmI and Poly(A) alleles in control women. Associations between VDR polymorphisms and breast cancer risk were estimated using unconditional logistic regression controlling for matching factors  and sampling weights. The BsmI BB and short Poly(A) SS genotypes were chosen as the respective referent categories based on published literature [The Pearson χterature ,16,19. Aterature . Known oVDR polymorphisms and breast cancer risk. In addition, given prior reports of the possible shared etiology of breast cancer and bone density, we assessed the potential modifying effects of a subset of breast cancer risk factors, specifically OC use among premenopausal women, HT use among postmenopausal women, smoking, and BMI, that are also known or suspected of having an effect on bone mineral density. Effect modification was explored in stratified models as well as in models using multiplicative interaction terms. In the multiplicative interaction models, a variable was considered an effect modifier if the P value for the interaction terms was less than or equal to 0.05. All models in which effect modification was explored were initially stratified by race, a matching variable, and menopausal status because of existing effect modification.We examined age, race, menopausal status and family history of breast cancer for potential modifying effects on the association between ® software . The study protocol was approved by institutional review boards at participating centers.Statistical analyses were performed using Stata/SEVDR polymorphisms . As expected by design, cases were similar to controls with regard to age and race, and the proportions with first-degree family history in our sampled group (37.4% cases and 23.6% controls) exceeded those in the underlying CARE study (17.0% cases and 9.7% controls). All subsequent analyses account for the sampled nature of the data, including the over-sampling of women with a first-degree family history of breast cancer, through sampling weight adjustment. For Poly(A), there was excellent observer agreement in the 6% of randomly selected duplicates of genotyping results that were included for quality control purposes (κ statistic of 0.99). Similar information was not collected for BsmI. Laboratory personnel were blinded to case-control status.Table Poly(A) genotypes in the controls . However, we did observe significant deviation for the BsmI genotype . BsmI and Poly(A) genotypes were in linkage disequilibrium for the Caucasian but not the African-American population .We did not observe deviation from the expected Hardy-Weinberg frequencies for the VDR polymorphism genotype frequencies in cases and controls (Table BsmI or Poly(A) and breast cancer risk for the combined study population  after controlling for sampling weights and matching factors: age, race, and study site. Similarly, we did not observe any associations between polymorphisms in BsmI or Poly(A) and breast cancer risk within strata of tumor stage, grade, histology, or receptor status (data not shown). After controlling for many known and suspected risk factors for breast cancer, we found no evidence that any of the variables mentioned above acted as confounding factors of these results. However, given the established variation in breast cancer risk factor profiles by menopausal status and because of the difference in the direction of the ORs by race, all further results are stratified by menopausal status and race  and, although not statistically significant, the risk in women carrying the heterozygous Bb genotype was also increased . A trend of increasing risk with increasing number of b alleles carried was statistically significant (P for trend = 0.03). We did not observe a significant association with breast cancer in any racial-menopausal status subgroup in relation to the Poly(A) polymorphism.The Poly(A) and BsmI genotypes among African-American women were not in linkage disequilibrium, we explored the combined effects of the Poly(A) and BsmI genotypes in this subpopulation (results not shown). An increased or decreased risk for breast cancer was not associated with the combination of the Poly(A) and BsmI genotypes. Stratifying or adjusting for age, menopausal status, or family history of breast cancer did not alter significantly the observed absence of association.Because the a priori hypothesized potential effect modifiers , as well as stratification by tumor characteristic  was not meaningful. Smoking was the only a priori hypothesized potential effect modifier that significantly modified the association between polymorphisms in BsmI or Poly(A) and breast cancer risk.Effect modification of the relationship of either genotype with risk for breast cancer by BsmI genotype and breast cancer risk among postmenopausal Caucasian women (P = 0.04). Among postmenopausal Caucasian women who reported never smoking, the risk for breast cancer in carriers of one or two b alleles was 2.2 times greater than for BB carriers , whereas among ever smokers (current or former) the risk for breast cancer was unrelated to BsmI genotype. Smoking status did not significantly modify the association between BsmI genotype and breast cancer risk among premenopausal Caucasian women (P = 0.21).Smoking status significantly modified the association between the P value for interaction between genotype and smoking status was significant (P = 0.04) whereas the P value for the interaction approached but did not achieve significance for postmenopausal African-American women (P = 0.08). Among postmenopausal African-American women who reported never smoking, the risk for breast cancer for carriers of one or two b alleles compared with individuals carrying the BsmI BB genotype was increased, albeit not statistically significantly , whereas among premenopausal African-American women who reported never smoking the risk for breast cancer in carriers of one or two b alleles was significantly decreased compared with BB carriers .Among premenopausal African-American women the Poly(A) genotype and breast cancer risk among postmenopausal Caucasian women (P = 0.02). Among postmenopausal Caucasian women who reported never smoking, the risk for breast cancer for carriers of any L allele was 2.3 times greater than that for carriers of the Poly(A) SS genotype , whereas among never smokers the risk for breast cancer was unrelated to Poly(A) genotype .Smoking status also significantly modified the association between In terms of the main effect of smoking, premenopausal Caucasian women who reported ever smoking had a statistically significant increased risk for breast cancer  as compared with women who reported never smoking. Smoking was not associated with an increased risk for breast cancer among postmenopausal Caucasian women, premenopausal African-American women, or postmenopausal African-American women.BsmI bb genotype. There was little evidence of a similar relationship in premenopausal or African-American women. In addition, there was no evidence of a relationship between the Poly(A) LL genotype and risk for breast cancer overall, or within any racial or menopausal status subgroup. We do note that genotype frequencies were different by menopausal status; however, there is no biologic explanation for this difference. Automated genotyping systems, to the degree that they are flawed, have a tendency to overcall homozygotes and undercall heterozygotes, but there is no reason to assume that bb or BB genotypes would be differentially miscalled according to any exposure, including menopausal status. This observation does not change the overall conclusions of the study, and the differing results for premenopausal and postmenopausal Caucasian women are probably accounted for by small numbers.Within the CARE study population, we observed a slight increase in risk for breast cancer among postmenopausal Caucasian women with the VDR and breast cancer risk have been limited in number and yielded inconsistent results. Three hospital-based case-control studies of Caucasian women in the UK reported an increased risk for breast cancer in carriers of BsmI bb genotype [BsmI bb genotype, and a nearly twofold increase in risk for carriers of the Poly(A) LL genotype, when controlling for age, HT, and menopausal status. Participants were matched for age at sampling (controls) versus diagnosis (cases), but they differed significantly in terms of menopausal status and HT use. The second study [BsmI bb genotype and Poly(A) LL genotype. The differences between this study and our own population-based study are significant, limiting our ability to compare results. The third study [BsmI bb compared with those with the BB genotype. The authors concluded that low circulating 25-hydroxyvitamin D levels, either alone or in combination with the BsmI Bb or bb genotype, may increase breast cancer risk.Prior studies of genotype ,10,16. Hgenotype . The larrd study  was distBsmI bb genotype or the Poly(A) LL genotype [VDR polymorphisms and breast cancer risk in a nested case-control study of largely Caucasian women aged 43 to 69 years within the Nurses' Health study. The authors genotyped 1,180 cases and 1,547 controls for the BsmI and 1,234 cases and 1,676 controls for the Fok1 polymorphisms. They observed a significantly increased risk for breast cancer among carriers of the ff genotype of Fok1  as compared with those with the FF genotype, but they found no association between the BsmI polymorphism and breast cancer risk  for BB versus bb. The conclusions did not change when the data were stratified by menopausal status. The Poly(A) variant was not tested.In contrast, several prior studies did not report an increased risk for breast cancer associated with either the genotype -20,28. Ogenotype , was conVDR polymorphism and breast cancer risk have included a substantial number of African-American women. It is important to note that we did not observe a statistically significant increased risk for breast cancer associated with either the BsmI bb genotype or the Poly(A) LL genotype among the African-American women in our study. The (nonsignificant) reduced risk for breast cancer associated with carrying either the BsmI Bb or bb genotypes among never smokers, observed in African-American women, was probably due to small numbers in the referent category and not the presence of the allele or the effect of not smoking. Significantly larger studies of African-American women would be needed to address that question more completely.To our knowledge, no prior studies of the VDR polymorphism that influence disease risk may be subject to environmental influences. For example, smoking is an inconsistent risk factor for breast cancer [VDR polymorphisms and breast cancer risk varied among postmenopausal Caucasian women who reported ever or never smoking. The effect modification results should, however, be interpreted with caution; although we did observe an association between breast cancer and the BsmI bb genotype among postmenopausal Caucasian women, we did not observe an association between breast cancer and smoking in the same subgroup.Genetic factors such as the t cancer , but thet cancer ,31. Alth3), a hormonal derivative of vitamin D3, has an antiproliferative and prodifferentiation effect on several cell types, specifically the squamous cells of the head and neck [VDR polymorphism (BsmI). Specifically, they observed a 57% reduction in risk among men with a plasma 25-hydroxyvitamin D level below the median who had the BB versus bb genotype . The risk reduction was strongest among older men .Vitamin D itself is proposed to have anticancer properties -34. Indeand neck . In a stand neck  reportedWe did not have information on vitamin D levels, via either dietary intake or metabolism from routine sun exposure. However, several studies -9 supporBsmI b allele in the development of breast cancer in women who never smoked, but not in those who did smoke. Why a subset of VDR genotypes would have a differential effect on smokers versus nonsmokers is unclear. One hypothesis is that there is some as yet undefined interaction between estrogen metabolism and smoking that is exacerbated by specific levels of vitamin D status.Our data support a role for the VDR polymorphisms and breast cancer risk. The population-based design, wider age range, and inclusion of both African-American and Caucasian women allow a more comprehensive portrayal of the frequency of mutations in the general population than has been available to date. Despite the generous sample size, the number of women with referent genotypes was small, resulting in an inability to make precise estimates. In addition, the study was limited by the fact that the gene was interrogated at only two positions at the 3' end of the gene. Newer technologies now permit a more comprehensive approach to genotyping, and variation within the entire gene will be readily examined in subsequent studies. The departure from Hardy-Weinberg equilibrium that we observed for the BsmI polymorphism could be attributed to any number of factors, including undetected ethnic diversity, biologic explanations, or genotyping misclassifications. We did not include blinded duplicates in the genotyping dataset, and therefore we did not have the opportunity to check for systematic genotyping errors.The strengths of the present study include its large size, generalizability, and the ability to control for most of the established risk factors for breast cancer. To date, this is the largest study of VDR gene to cancer risk.One difficulty in studies of single nucleotide polymorphisms, such as this, is the issue of whether the variant being tested is truly associated with the disease or whether there is another polymorphism in linkage disequilibrium with the polymorphism under consideration that could be responsible for the observed association. Only when the whole gene can be examined along with associated regulatory elements can we develop a clearer picture of the underlying biology that consistently links the BsmI polymorphism at the 3' end of the VDR gene in postmenopausal Caucasian women and sheds light on the differential risk observed in menopausal status and race. Contributions of vitamin D intake to risk for breast cancer and bone mineral density, as well as potential effect modification by history of smoking, should be further studied to elucidate the precise biologic mechanisms.This study provides additional support for an increased risk for breast cancer associated with the BMI = body mass index; bp = base pairs; CARE = Contraceptive and Reproductive Experiences; HT = hormone therapy; OC = oral contraceptive; OR = odds ratio; PCR = polymerase chain reaction.The authors declare that they have no competing interests.BT performed the statistical analysis and drafted the manuscript. KM participated in study design and coordination, hypothesis generation, and manuscript preparation and also oversaw data collection and statistical analyses. JD, LB, PM, and BS participated in study design, data collection, and manuscript review. DD participated in the design of the study, manuscript review, and performed statistical analyses related to sampling. GU participated in manuscript review. MH carried out molecular genetic analyses. EO oversaw molecular genetic analyses, participated in study design, hypothesis generation, and manuscript preparation. All authors read and approved the final manuscript.


N=1197), were identified using the Saskatchewan Cancer Agency cancer registry. Women ⩾50 years of age, eligible for coverage by the Saskatchewan Prescription Drug Plan, were included in the study. Four controls per case were age matched to cases, using incidence density sampling. The outpatient prescription drug plan database was used to ascertain oestrogen prescriptions. Women were classified according to history of transdermal (TDE) and oral (OE) oestrogen use. Conditional logistic regression was used to estimate odds ratios (ORs) and 95% confidence intervals (CIs). Compared with women who had never used HRT, ORs for <3 and ⩾3 years of TDE use and colorectal cancer were 0.69 (95% CI: 0.43–1.10) and 0.33 (95% CI: 0.12–0.95), and for OE use were 0.90 (95% CI: 0.73–1.01) and 0.75 (95% CI: 0.60–0.93), respectively. The risk reduction for colorectal cancer with TDE may be greater in magnitude than that which has been reported for oral HRT.The aim of this study was to examine the effects of oral and transdermal oestrogen replacement therapy on the risk of colorectal cancer. Data from a nested case–control study, designed to investigate the effect of hormone replacement therapy (HRT) on colorectal cancer were analysed. New cases of colorectal cancer, diagnosed between 1992 and mid-1998 ( The release of results from the Women's Health Initiative (WHI)  and the The results from the WHI  randomisvs transdermal) to clinically important outcomes. Faced with the decision to begin HRT or not, women need to weigh the health risks against the benefits of menopausal symptom control. Provided that there is sufficient knowledge about the health effects of various HRT formulations and modes of delivery, women and their physicians could potentially individualise treatment with HRT to minimise the health risks and optimise benefit.While evidence from clinical studies demonstrate important metabolic and biologic differences between oral and transdermal HRT  is a provincial government department that maintains electronic health service databases as part of its publicly funded health system. These databases were used as sources of health services information for this study. Over 99% of the Saskatchewan population are registered with Saskatchewan Health and about 91% are eligible for outpatient prescription drug benefits  cancer registry, which has been in existence since 1932 and computerised since 1970 . Cancer in situ) prior to the index date. Saskatchewan Cancer Agency records were used to verify that participants had not had previous cancer diagnoses.All cases and controls eligible for this study had to have had at least 5 years of registration with SH, had to have been eligible for outpatient prescription drug benefits and not have been diagnosed with cancer  to each identified case. Specifically, a control had to have been living in Saskatchewan at the time the colorectal cancer case (to whom she was matched) was diagnosed with cancer (index date). For each case, 16 potential controls were randomly sampled using incidence density sampling , with reDispensing data were ascertained for all forms of oestrogen: oral, transdermal patch, and vaginal creams and rings listed in the Saskatchewan Formulary. For each woman in the study, the type of formulation, strength, and number of oestrogen units dispensed, prior to index dates, were obtained from the records of the outpatient prescription drug plan database, originally established in 1975. Prescription drug dispensing data were ascertained for each woman from the time her prescription drug coverage began, until the date of colorectal cancer diagnosis for cases or assigned index dates for controls.Definitions of oestrogen exposure). There was also a break in complete data capture for all dispensed prescriptions between July 1, 1987 and December 31, 1988. For this period of time, the HRT exposure assigned to each woman was based on her exposure status before and after the 18-month period. No exposure was assigned if she did not have a prescription dispensed during the year before or after the break, one additional year of transdermal or OE exposure was assigned if she had had at least one transdermal or OE prescription dispensed during the year prior to, or, during the year following the break.Transdermal oestrogen replacement was listed in the Formulary on July 1, 1988 with unrestricted coverage, and it was moved to restricted coverage on January 1, 1997  or long-term use (⩾3 years). A 3-year threshold was selected, The use of oral contraceptives, cardiovascular system (CVS) drugs, central nervous system (CNS) drugs, and prescription nonsteroidal anti-inflammatory drugs (NSAIDs) and vitamins, and other hormones were also ascertained from the outpatient prescription drug plan database for the period of time preceding index dates. The history of having had a sigmoidoscopy and the frequency of physician visits during the 5-year period prior to index dates were ascertained for all cases and controls, using records of the medical services database.a priori, were assessed using the criteria of observing a change in ORs of 10% or more. Backward and forward model selection methods were used to identify the best regression model.Conditional logistic regression was used to calculate the odds ratios (ORs) and 95% confidence intervals (CIs). The effect of age was controlled for with matching and, where matching was broken, age-adjusted ORs were obtained with the inclusion of age in the regression model. The confounding effects of covariates, identified All associations between TDE and OE and risk of colorectal cancer were examined among all women and for the subgroup who had not had a sigmoidoscopy 3–5 years prior to index dates. The analysis could not be carried out among women who had had a sigmoidoscopy during this period of time, because too few women had had the procedure.An unmatched analysis was also carried out where a direct comparison was made between TDE and OE on the risk of colorectal cancer. In this analysis, women exposed to TDE were compared with women who had used OE exclusively (reference group). Age-adjusted ORs and 95% CIs were calculated using unconditional logistic regression.A statistical test for linear trend was carried out for the duration of HRT use by redefining short and longer duration (<3 or ⩾3 years) of TDE and OE use as ordinal variables, and assessing their level of significance in a logistic regression model. All statistical analyses were carried out using SAS software .vs 2.3% for cases and controls, respectively). Only minor differences were observed for having had a sigmoidoscopy during the 2–5 and 3–5 years preceding the index dates. During the year preceding the index date, 65.1% of the cases also visited their physician 15 times or more, compared with 28.8% of controls. The frequency of physician visits 2–5 years prior to index dates did not differ greatly between cases and controls.In all, 1261 cases and 4916 age-matched controls had index dates between January 1, 1992 and mid-1998. Of these, 1197 cases had histologically confirmed colorectal cancer. The health-related characteristics of these 1197 colorectal cancer cases and their 4669 controls are outlined in The number of women, among cases and controls, exposed to various oestrogen formulations prior to the reference date, are presented in Prescription drug use only minimally altered ORs for oestrogen and colorectal cancer (less than 10%), and therefore only age-adjusted ORs are presented. ‘Ever’ use of OE was associated with an OR of 0.82 (95% CI: 0.70–0.97) . Similar point estimates were obtained for women who had not had a sigmoidoscopy 3–5 years prior to index dates. The test for linear trend remained significant for all OE use (P=0.02), but not for OE only use (P=0.07).The association between colorectal cancer risk and duration of OE use is outlined in P for trend=0.01). Similar results were observed for TDE use among women who had not had a sigmoidoscopy 3–5 years prior to index dates, with the test for linear trend remaining significant (P=0.01). For TDE users only, the same trend in decreasing ORs was observed with increasing duration of use; however, only five cases had been exposed to TDE only and only one case was exposed for 3 or more years.For TDE use, among women of whom some had used OE in the past, the ORs were 0.69 (95% CI: 0.43–1.10) for less than 3 years of TDE use, and 0.33 (95% CI: 0.12–0.95) for 3 or more years of TDE use   is avoided with TDE . The estIn summary, the reduction in colorectal cancer risk with TDE use has not previously been reported and warrants further investigation. Knowing more about the differences in health outcomes between TDE and OE could help women and their physicians individualise therapy in order to minimise the risks and maximise the benefits of treating distressing menopausal symptoms.


A number of breast cancer risk prediction models have been developed to provide insight into a woman's individual breast cancer risk. Although circulating levels of estradiol in postmenopausal women predict subsequent breast cancer risk, whether the addition of estradiol levels adds significantly to a model's predictive power has not previously been evaluated.Using linear regression, the authors developed an imputed estradiol score using measured estradiol levels (the outcome) and both case status and risk factor data  from a nested case-control study conducted within a large prospective cohort study and used multiple imputation methods to develop an overall risk model including both risk factor data from the main cohort and estradiol levels from the nested case-control study.P < 0.001) after the addition of imputed estradiol.The authors evaluated the addition of imputed estradiol level to the previously published Rosner and Colditz log-incidence model for breast cancer risk prediction within the larger Nurses' Health Study cohort. The follow-up was from 1980 to 2000; during this time, 1,559 invasive estrogen receptor-positive breast cancer cases were confirmed. The addition of imputed estradiol levels significantly improved risk prediction; the age-specific concordance statistic increased from 0.635 ± 0.007 to 0.645 ± 0.007 (Circulating estradiol levels in postmenopausal women appear to add to other lifestyle factors in predicting a woman's individual risk of breast cancer. Breast cancer risk prediction models have been developed for use as an entry criterion into breast cancer chemoprevention trials , in counseling women on the potential use of chemopreventives, and to provide insight into a woman's individual breast cancer risk -4. The iIn postmenopausal women, circulating levels of estradiol predict subsequent breast cancer risk -10, partThe Nurses' Health Study cohort was established in 1976, when 121,700 female US nurses (30 to 55 years old) responded to a mailed questionnaire that inquired about reproductive history and a range of lifestyle factors in addition to disease diagnoses ,12. Foll+ cases only, as we previously observed that plasma estradiol most strongly predicted this tumor subtype . The incidence rate of breast cancer is assumed to be approximately proportional to the number of precancerous cells.The general rationale for a log-incidence model is that the number of precancerous cells increases multiplicatively with time but that historical exposures differentially affect the rate of increase. Specifically, for breast cancer, the number of precancerous cells is assumed to increase annually at the rate of exp(β0) = RR for a 1-year increase in age at menarche among nulliparous women, exp [-(β0 + β2)] = RR for a 1-year increase in age at menarche among parous women, and so forth. In this analysis, women were followed until they had an event (ER+ breast cancer) or were censored if they developed (a) ER- breast cancer, (b) breast cancer in which ER status is unknown, or (c) other types of cancer except nonmelanoma skin cancer or (d) if they died.The log-incidence model was fit using iteratively reweighted least squares with PROC NLIN in SAS  (1996). The parameters of the model are readily interpretable in a relative risk (RR) context. For example, exp given under the rare disease assumption by:From the main study, we can obtain Pr, where under the rare disease assumptionWe want to estimate Pr as a continuous variable, y = 1 if case and 0 if control, and Zimp = a subset of the other covariates Z in the risk prediction model. Zimp was ascertained by first forcing in y and then using stepwise-up regression to determine the subset of components of Z in the main study which were significantly associated with x at the 5% level.where y as a covariate in Equation 6 is to account for this relationship in the main study as well. In addition, because there is substantial overlap between the estradiol distribution of cases and controls, we used an imputation strategy to estimate x by adding error to the prediction such that for each main study participant we obtainIn the blood study, estradiol levels on average were higher for cases than controls. The rationale for including ei = an error term that is normally distributed with mean 0 and variance σ2, (b) yi = 1 if a breast cancer case and = 0 otherwise, (c) σ2 is estimated from Equation 6, and e is obtained by the RANNOR function of SAS so as to add error to the estimate of x for individual women. We then fit the model in Equation 5 using x, thus obtaining the modelwhere (a) Since the parameter estimates in Equation 8 may be influenced by the random error introduced in Equation 7, we repeated this imputation approach four additional times and used multiple imputation  to combiij = the number of breast cancer cases, Nij = the number of person-years, and pij = Xij/Nij, which is the estimated incidence rate within the ith age-specific risk decile for model A and the jth age-specific risk decile for model B, and let ln(pij) = αi + β(j - 1). 100% ×  and 0, respectively. The RRs for the parous versus nulliparous woman are exp [-0.0030(102)] = 0.74 (95% CI = 0.64 to 0.85) without and exp [-0.0020(102)] = 0.82 (95% CI = 0.69 to 0.96) with controlling for loge estradiol. Finally, the regression coefficient for both the duration after natural menopause and the duration after bilateral oophorectomy increased after adjusting for loge estradiol. To interpret this finding, we compare two postmenopausal women with age at natural menopause of 45 and 55 years, respectively. The RRs of breast cancer for the second compared with the first woman are exp [10(0.102-0.049)] = 1.7 (95% CI = 1.4 to 2.0) without and exp [10(0.101-0.056)] = 1.6 (95% CI = 1.3 to 1.9) with adjusting for loge estradiol. It appears that part of the effect of BMI, parity, and late menopause on the incidence of breast cancer is mediated in part by changes in loge estradiol caused by obesity (increase), multiparity with first birth at an early age (decrease), and delayed menopause (increase), respectively.In Table P < 0.001), indicating that there is a significant estimated 67% increase in breast cancer incidence for an increase of one model B age-specific risk decile, holding the age-specific model A risk decile constant. This indicated that there is substantial increased predictive power upon adding loge estradiol to the risk prediction model.We now present the cross-classification of model A × model B risk decile in Table P < 0.001). Using our secondary approach (applying the population-based risk scores to the subset of women in the nested case-control study), the RRs of breast cancer by plasma hormone level were similar, though slightly attenuated, compared with those in Table + breast cancer with increasing quartile of estradiol were 1.0, 1.5, 1.4, and 2.5 (95% CI = 1.5 to 4.2). Finally, in Table 5(0.4%) for women in the first decile and 2,982/105 (3.0%) for women in the 10th decile (RR = 6.8), indicating substantial differences in absolute risk according to the model B risk equation.In addition, we compared the age-specific C statistics between model A versus model B. We found C statistics of 0.635 ± 0.007 for model A and 0.645 ± 0.007 for model B . Thus, it was not possible within this small data set to provide a sufficiently precise evaluation of the Rosner and Colditz model (which contains 22 beta coefficients); in our initial attempt to evaluate the model, all beta coefficients had wide CIs. In our secondary analyses, in which we used the risk score within the nested case-control data set, plasma estradiol again contributed significantly to the model. However, the RR for estradiol in the secondary analysis for Q4 versus Q1 was 2.5 (95% CI = 1.5 to 4.2) when controlling for other risk factors using the Rosner-Colditz risk scores versus 3.3 (95% CI = 1.8 to 6.0) when individual risk factors were used within the case-control study. One would expect that the effects of other risk factors are more accurately measured by a single risk score derived from a large cohort study than individual risk factors derived from a relatively small nested case-control data set. More generally, this may indicate better control for confounding in small case-control studies based on risk scores derived from large cohorts versus internal control for confounding based on individual risk factors whose regression coefficients are poorly estimated in small case-control studies. With further follow-up and within a collaboration across several cohorts at other institutions, we plan to re-evaluate the case-control approach.e (estradiol). The correlation between BMI and estradiol was expected given that aromatization of androgens to estrogens in postmenopausal women occurs in adipose tissue [In imputing estradiol levels, only BMI, the birth index, and duration of postmenopause (in addition to case status) were significant predictors of loge tissue . The asse tissue -28. The e tissue , did note tissue . To our With the inclusion of imputed estradiol, the C statistic increased from 0.635 to 0.645, which is a modest improvement but suggests reasonable discriminatory ability overall. However, the reclassification table approach (Table In summary, our data indicate that circulating estradiol levels in postmenopausal women may contribute significantly to current risk prediction models. Further assessment of estradiol in other studies and of other biomarkers that predict risk is needed to continue to improve our ability to predict breast cancer risk and inform prevention strategies. Similar approaches can be used to incorporate other breast cancer biomarkers in overall risk prediction models.BBD = benign breast disease; BMI = body mass index; C statistic = concordance statistic; CI = confidence interval; ER = estrogen receptor; PMH = postmenopausal hormone; Q1, Q2, Q3, and Q4 = first, second, third, and fourth quartile; RR = relative risk.The authors declare that they have no competing interests.BR, GAC, and SEH contributed to the design of the study and to the analysis and interpretation of the data. JDI contributed to the revision of the manuscript and added important clinical insight. All authors read and approved the final manuscript.


Although pregnancy-related factors such as nulliparity and late age at first full-term pregnancy are well-established risk factors for invasive breast cancer, the roles of these factors in the natural history of breast cancer development remain unclear.in situ (CIS) and 2,828 with invasive breast cancer between 1995 and 2007. Multivariable Cox proportional hazards regression methods were used to estimate relative risks associated with parity, age at first full-term pregnancy, breastfeeding, nausea or vomiting during pregnancy, and preeclampsia.Among 52,464 postmenopausal women participating in the California Teachers Study (CTS), 624 were diagnosed with breast carcinoma P trend < 0.01). Women having four or more full-term pregnancies had a 31% lower breast CIS risk  and 18% lower invasive breast cancer risk . Parous women whose first full-term pregnancy occurred at age 35 years or later had a 118% greater risk for breast CIS  and 27% greater risk for invasive breast cancer  than those whose first full-term pregnancy occurred before age 21 years. Furthermore, parity was negatively associated with the risk of estrogen receptor-positive (ER+) or ER+/progesterone receptor-positive (PR+) while age at first full-term pregnancy was positively associated with the risk of ER+ or ER+/PR+ invasive breast cancer. Neither of these factors was statistically significantly associated with the risk of ER-negative (ER-) or ER-/PR- invasive breast cancer, tests for heterogeneity between subtypes did not reach statistical significance. No clear associations were detected for other pregnancy-related factors.Compared with never-pregnant women, an increasing number of full-term pregnancies was associated with greater risk reduction for both breast CIS and invasive breast cancer (both in situ versus invasive, and hormone receptor-positive versus -negative breast cancer merits further exploration.These results provide some epidemiologic evidence that parity and age at first full-term pregnancy are involved in the development of breast cancer among postmenopausal women. The role of these factors in risk of  The clarification of these issues may shed light on a deeper understanding of the roles of pregnancy-related factors in the natural history of breast cancer development.Although pregnancy-related factors such as nulliparity and late age at first full-term pregnancy are well-established risk factors for invasive breast cancer, it remains unclear whether such factors have similar effects on breast carcinoma in situ (DCIS), lobular carcinoma in situ (LCIS), and other relatively rare forms of breast CIS, with DCIS being the predominant subtype. The term 'in situ' indicates that neoplastic cells are present but have not spread past the boundaries of ducts or lobules where the tumor initially developed. Estimates of the percentage of patients with DCIS that will progress to invasive disease over a period of 10 or more years vary widely, ranging from 14 to 53% [in situ breast cancer [Breast CIS includes ductal carcinoma t cancer ,3. In coGiven that breast CIS is a potential precursor to invasive disease, one would expect that breast CIS and invasive breast cancer would share risk factors involved in the early stages of breast cancer development while factors affecting disease progression would be associated only with invasive breast cancer risk. Few studies have made direct comparisons of the consistency of risk estimates for pregnancy-related risk factors between breast CIS and invasive breast cancer within a study population, particularly among women whose reproductive years have ended. In studies that provided results for postmenopausal women, parity was associated with a decreased risk for both postmenopausal breast CIS and postmenopausal invasive breast cancer; results for age at first full-term pregnancy were inconsistent -8. None Parity, early age at first full-term pregnancy, and breastfeeding influence the risk of breast cancer predominantly through hormonal mechanisms that involve estrogen and progesterone ,10. PrevBoth a systematic review  and a meThus, important questions remain regarding the roles of pregnancy-related factors in the development of breast CIS and invasive breast cancer among women who have completed their reproductive years. The authors examined the associations of parity, age at first full-term pregnancy, duration of breastfeeding, nausea or vomiting during pregnancy, and preeclampsia with breast CIS and invasive breast cancer  among postmenopausal women participating in the prospective California Teachers Study (CTS) cohort.Details of the CTS have been described previously . BrieflyAll information on pregnancy-related factors except nausea or vomiting during pregnancy and preeclampsia was collected in the questionnaire completed by participants when they joined the cohort. Participants reported information on race/ethnicity, family history of breast cancer, age at menarche, detailed pregnancy (including ages at and outcomes for each pregnancy) and breastfeeding histories, menopausal status, use of menopausal hormone therapy, height, and weight. Participants reported information regarding nausea or vomiting during pregnancy and preeclampsia in the second questionnaire completed in 1997-1998. Preeclampsia was defined as a condition that can occur during the second half of a pregnancy and is marked by elevated blood pressure, protein in the urine and fluid retention.For the current analysis, using information provided when the cohort was formed, 8,867 women who were not California residents were excluded, followed by 6,350 women who had a prior history of breast CIS or invasive breast cancer or an unknown cancer history and by 5,107 women who were 80 years or older. This yielded a preliminary analytic cohort of 113,155 women. In order to ensure that each woman's complete pregnancy history was covered, eligibility for this analysis was restricted to women who were postmenopausal at cohort entry . Menopausal status was determined after reviewing a woman's age at last menstrual period, reason for cessation of menstrual periods, hysterectomy status, oophorectomy status and current use of hormonal therapy. Women were considered to be postmenopausal if they reported a natural menopause more than six months before completing the baseline questionnaire n = 31,207), if they had a bilateral oophorectomy , if they were 56 years or older and not menstruating regularly , or if their periods had stopped due to other reasons including pituitary adenoma, medication, chemotherapy, radiation treatment, or another reason . The age criterion was based on previous work indicating that among those with natural menopause 97% were postmenopausal by age 56 years [,207, if Person-time of follow-up for each woman began with the date she completed the baseline questionnaire and ended with the first of the following events: a breast cancer diagnosis , a permanent move outside of California , death , or December 31, 2007 .The analytic cohort included 8,937 never-pregnant and 43,527 ever-pregnant women. Of the 43,527 ever-pregnant women, 371 experienced pregnancy but did not provide outcome, and 1,957 had only incomplete pregnancies. Women who experienced pregnancy and did not provide an outcome and those who had only incomplete pregnancies were included when testing the effect of any pregnancy (ever vs. never), but were excluded from analyses of parous women. The effect of incomplete pregnancy on breast cancer risk has previously been studied in this cohort .The 41,199 women who had at least one stillbirth or live childbirth were referred to as parous throughout this article. Still and live births were combined to obtain the total number of full-term pregnancies. Parous women who were missing information on breastfeeding or age at first full-term pregnancy (n = 117) were additionally excluded from analyses restricted to parous women. Analyses of parous women were based on 41,082 women of whom 479 were diagnosed with breast CIS and 2,193 were diagnosed with invasive breast cancer during follow-up.Of 43,527 ever-pregnant postmenopausal women at cohort entry, 32,084  provided information on nausea or vomiting during any pregnancy and 31,459  provided information on preeclampsia in the CTS' second questionnaire, which was mailed in 1997, and were included in the analysis of these factors.Since 1990, National Cancer Institute designated SEER Program registries, including those comprising the CCR, routinely abstract ER and PR status from the medical records following a breast cancer diagnosis. Among 2,828 invasive breast cancers diagnosed in the current analytic cohort, ER status was available for 2,443 . Both ER status and PR status were available for 2,326 .The use of CTS participants' data for this analysis was approved by the Institutional Review Boards at the City of Hope, the University of Southern California, the Cancer Prevention Institute of California , and the University of California at Irvine, and by the Committee for the Protection of Human Subjects, California Health and Human Services Agency.The relative hazard (represented as the relative risk (RR) and its 95% confidence interval (CI)) for the association between pregnancy-related factors and breast cancer risk was estimated using multivariable Cox proportional hazards regression methods . Separatin situ disease stage undetected. For analyses of invasive breast cancer, women who developed breast CIS were censored on the date of this diagnosis. For analyses of each receptor subtype of invasive breast cancer, women who developed all other receptor subtypes or had an unknown subtype were censored on the date of this diagnosis.For all analyses of breast CIS, women diagnosed with invasive breast cancer during follow-up were excluded as the diagnosis of invasive disease presumes they have passed through the a priori, as potential confounders: race , family history of breast cancer in a first degree relative, that is, mother, father, sister or brother , age at menarche , menopausal hormone therapy (HT) use , only estrogen in combination with progesterone therapy (E+P), both types of HT, unknown type of HT) and body mass index .In the Cox regression models, the time scale was defined by age (in days) at cohort entry and age (in days) at event or censoring (exit). All multivariable models were adjusted for the following factors, selected Multivariable models for parous women were additionally mutually adjusted for number of full-term pregnancies , age at first full-term pregnancy , and duration of breastfeeding .The associations of breast CIS or invasive breast cancer with nausea or vomiting during pregnancy or preeclampsia were explored among postmenopausal women with at least one pregnancy as reported at cohort entry who had complete information on one or both of these variables in the second mailed questionnaire. Nausea or vomiting during any pregnancy was examined as was the number of pregnancies during which the ever-pregnant woman experienced these symptoms  and whether treatment was received for these symptoms during the most recent pregnancy or during other pregnancies. Preeclampsia was evaluated as never occurring during a pregnancy, occurring during the most recent pregnancy, or not occurring during the most recent pregnancy but having occurred during at least one of other pregnancies. Multivariable models for nausea or vomiting during pregnancy or preeclampsia were additionally adjusted for pregnancy history .Tests for trend were conducted to examine the dose-response relationship of breast CIS or invasive breast cancer with number full-term pregnancies, age at first full-term pregnancy, duration of breastfeeding, and number of pregnancies during which participant experienced nausea or vomiting by fitting ordinal values corresponding to each exposure category in the statistical models and determining whether the slope parameter differed from zero using the Wald test .Tests for homogeneity  of RR estimates were performed  by constructing a Z test of the differences in log RR divided by the square root of the sum of the variances of the two log RR estimates for the following comparisons: breast CIS vs. invasive breast cancer, ER+ vs. ER-, and ER+/PR+ vs. ER-/PR- breast cancer.The analyses for breast CIS were repeated among 44,671 (90%) postmenopausal women who reported having their most recent mammogram less than three years before the CTS baseline survey. The results remained the same. The analysis for breast CIS was also repeated after excluding 51 women diagnosed with LCIS. Again, the results remained essentially identical. The results for breast CIS including LCIS from all postmenopausal women without invasive breast cancer  are presented.P-values are reported for tests for trend and for homogeneity of RR estimates. All statistical analyses were performed using SAS version 9.1 .Two-sided Women in the analytic cohort were followed for an average of 10.5 years. Table P trend = 0.008) and invasive breast cancer (P trend = 0.003).Compared with never-pregnant women, women who experienced any pregnancy had reduced risks of breast CIS  and invasive breast cancer  Table . CompareP trend = 0.09; invasive breast cancer, P trend = 0.002) . In addition, these risk estimates for both breast CIS and invasive breast cancer did not vary within subgroups defined by body mass index , HT use , and race .Further, the risk estimates for parity, age at first full-term pregnancy, and breastfeeding did not differ statistically between breast CIS and invasive breast cancer  or if the woman received treatment for nausea or vomiting of pregnancy (a measure of severity of the condition) or the timing of that treatment (for the most recent or only for earlier pregnancies) , but was not associated with invasive breast cancer  with hyperemesis gravidarum [Since hyperemesis gravidarum, that is, severe nausea combined with persistent vomiting during pregnancy, has been linked to an elevated serum estradiol level , one migavidarum . In contavidarum  while anavidarum . Both ofPreeclampsia has been associated with lower maternal serum levels of estriol  and insuStrengths of the current study include its size, the large number of women who have been diagnosed with an incident breast cancer, the ability to identify and confirm cancer diagnoses through the CCR, California's high-quality statewide cancer registry, and collection of pregnancy-related information prior to the diagnosis of breast CIS or invasive breast cancer.expert laboratory found only small differences in risk estimates for ER+/PR+ and ER-/PR- breast cancer between the two sources [P > 0.07). These pregnancy-related factors were also similar between women having invasive breast cancer who had information on both receptor status markers and those without this information, except that those with the information for both ER and PR status were slightly more likely to have never been pregnant . This difference could cause an overestimate of any protective effect of pregnancies on all subtypes, but it is unlikely that this bias would be restricted to ER+/PR+ invasive breast cancer. In addition, the relative risk estimates for ER+/PR+ subtype differed minimally from those for ER+ subtype and the estimates for ER-/PR- subtype were similar to those for ER- subtype. Therefore, the missing information on ER and/or PR status is unlikely to have introduced measurable bias in this study. Since 75% of women with breast CIS had no data for ER/PR status, we were unable to examine the association between pregnancy-related factors and the risk of breast CIS by ER/PR status.Several limitations of the current study must be considered. ER and PR status results were collected by the regional registries in California from pathology laboratories located throughout the state; these laboratories may vary in their application of immunohistochemical methods and the cutpoints used to assign a positive status. However, it is unlikely that any methodological differences would influence the observed associations, as a large validation study comparing registry reports of receptor status to those of a single  sources . Althoug sources ,61. WomeAnother limitation is that the analyses for nausea or vomiting during pregnancy and preeclampsia were based on approximately 73% of ever-pregnant postmenopausal women. These women were similar to those without information on these two factors in terms of race, family history of breast cancer in a first degree relative, age at menarche, HT use, body mass index, number of full-term pregnancy, age at first full-term pregnancy, and duration of breastfeeding. Therefore, selection bias is unlikely to be an important explanation for the results for these two factors.These results provide some epidemiologic evidence that parity and age at first full-term pregnancy are involved in the development of breast cancer among postmenopausal women. This study did not provide evidence of any association of breastfeeding with postmenopausal diagnosis of breast CIS or invasive breast cancer overall or by hormone receptor subtypes. Further, nausea or vomiting during pregnancy was not associated with stage of breast cancer and no evidence was observed that preeclampsia is clearly associated with lower breast cancer risk.in situ; CTS: California Teachers Study; DCIS: ductal carcinoma in situ; E: estrogen; ER: estrogen receptor; HT: hormonal therapy; LCIS: lobular carcinoma in situ; P: progestin; PR: progesterone receptor; RR: relative risk; SEER: Surveillance, Epidemiology, and End Results.BMI: body mass index; CCR: California Cancer Registry; CI: confidence interval; CIS: carcinoma The authors declare that they have no competing interests.LB and PLHR participated in the study design and supervised the data collection. Data management was conducted by HM, JSH, and SM. HM and SM conducted data analyses under the supervision of LB. HM drafted the manuscript. All authors participated in the revision of the manuscript and have read and approved the final version.


To date, few epidemiologic studies have examined the relationship between environmental PCDD/F exposure and breast cancer in human populations. Dioxin emissions from municipal solid waste incinerators (MSWIs) are one of the major sources of environmental dioxins and are therefore an exposure source of public concern. The purpose of this study was to examine the association between dioxins emitted from a polluting MSWI and invasive breast cancer risk among women residing in the area under direct influence of the facility.We compared 434 incident cases of invasive breast cancer diagnosed between 1996 and 2002, and 2170 controls randomly selected from the 1999 population census. A validated dispersion model was used as a proxy for dioxin exposure, yielding four exposure categories. The latter were linked to individual places of residence, using Geographic Information System technology.The age distribution at diagnosis for all cases combined showed a bimodal pattern with incidence peaks near 50 and 70 years old. This prompted us to run models separately for women aged 20–59 years, and women aged 60 years or older. Among women younger than 60 years old, no increased or decreased risk was found for any dioxin exposure category. Conversely, women over 60 years old living in the highest exposed zone were 0.31 time less likely  to develop invasive breast cancer.Before speculating that this decreased risk reflects a dioxin anti-estrogenic activity with greater effect on late-onset acquired breast cancer, some residual confounding must be envisaged. Established risk factors for breast cancer are hormonally mediated: age, family history of breast cancer, early menarche, late menopause, late first full-term pregnancy or nulliparity, breast density, and benign breast disease ,2. But tp-dioxins (PCDD) and 135 polychlorinated dibenzofurans (PCDF). Seventeen tetrachloro-substituted congeners are toxic, with 2,3,7,8-tetrachlorodibenzo-p-dioxin (TCDD) being the most potent. The U.S. Environmental Protection Agency and the International Agency for Research on Cancer have classified TCDD as a human carcinogen [Dioxin is the name given to two classes of organochlorine compounds; 75 polychlorinated dibenzo-rcinogen ,5.This compound is also known to disrupt multiendocrine pathways in animals at body burdens which are close to those present in the background human population . TCDD elHowever, accumulating evidence suggests that TCDD also possesses estrogen-like activity. TCCD induces an estrogen-like gene expression profile in the uteri of immature ovariectomized mice in the absence of histopathological or morphological manifestations , and medThe opposing actions of TCDD, anti-estrogenic in the presence of estrogen and estrogenic in its absence, suggest that the effects of TCDD may vary depending on developmental stage at exposure . The abi10 TCDD levels was significantly increased by 2.1-fold , while adjusting for established breast cancer risk factors.To date, few epidemiologic studies have examined the relationship between PCDD/Fs and breast cancer in human populations. Most of these have consisted of occupational cohorts yielding conflicting results . Warner Dioxin emissions from municipal solid waste incinerators (MSWIs) are one of the major sources of environmental dioxins and are therefore an exposure source of public concern. Our team recently detected a cluster of non-Hodgkin lymphoma (NHL) in an area that contains a MSWI with high dioxin emission levels  . We subsBoth the suggestive results by Warner et al. , and the3, whereas the European guide value is 0.1 ng I-TEQ/m3. With such conditions, it could be expected that in the last two decades remarkable amounts of PCDD/Fs were released into the environment.The MSWI under investigation has been fully described elsewhere . SituateSince a dispersion model provided a reliable proxy for dioxin exposure on the northeast side of the MSWI (simple terrain), but not in the southwest direction (complex terrain) , the 90 Invasive breast cancer incidence data were provided by the Doubs cancer registry. The latter was established in 1977 and is complete for breast cancer cases, as ascertained by the ratio of the number of deaths to the number of cases registered during 1993–1997, which at 32% (for the Doubs region) is very similar to those reported in other Western countries [We selected female controls from a reliable and accessible database, the 1999 population census. Because of French privacy laws and confidentiality requirements the only individual (and anonymous) data available to researchers are sex, age categories , and residence in a given block. The block is the smallest level of geographic resolution in the French census database and is defined only in densely populated areas. Each block is typically a quadrangle bounded by four streets. In the 1999 census, there are 590 blocks in northeast Besançon , aggregated in 42 block groups . Unfortunately, no other risk factor data at the individual or block level are available to researchers. However, a set of social and demographic information is available for each block group. We randomly selected age-matched population-based controls, according to a 5-to-1 matching procedure.Dioxin exposure assessment has been fully described elsewhere . In no iThe respective contours of these modeled ground-level air concentrations  were digitized and contoured onto the surface of the map with geographic information system (GIS) tools. We then overlaid a map of blocks onto the digital dioxin concentration map to attribute a dioxin concentration category to each of the 590 northeast blocks  Figure . From thWe used conditional logistic regressions to calculate odds ratios (ORs) and 95% CIs for each level of dioxin exposure estimated from the dispersion model. A set of dummy variables was generated for this categorical scale variable using the lowest category as the reference group. Models were run with LogXact software .From 1996 to 2002, a total of 434 invasive breast carcinomas was diagnosed in the northeast side of the city of Besançon, and 2170 population controls were therefore randomly selected.These cases corresponded to an age-standardized (world) incidence rate of 81.4 per 100,000, to be compared to the age-standardized (world) incidence rate of 90.4 per 100,000 for France, as estimated in 2000 , and the age-standardized (world) incidence rate of 76.9 per 100,000 for the Doubs region that comprises the city of Besançon .All but 12 cases were histologically confirmed (97.23%). Duct NOS and lobular carcinomas were the most common histologic subtypes accounting for 79.26% and 11.75%, respectively.The age distribution at diagnosis for all breast cases combined showed a bimodal pattern with incidence peaks (or modes) near ages 50 and 70 years Figure . This prThe distributions of these cancer patients by age bands and dioxin exposure categories are displayed in Table Excluding the 12 non-histologically confirmed cases did not alter the findings (results not shown).Overall, our results do not provide evidence for an association between dioxins emitted from the MSWI and breast cancer risk among younger women, but show a decreased risk among older women residing in areas classified as highly exposed to dioxins.Our study has a number of strengths. It is population-based and involves cases actively identified through multiple sources within a defined geographic area. Controls were randomly selected from a complete directory of people residing in the same area with a modest but relevant and reliable list of characteristics available at low cost (census data). The 5-to-1 matching procedure , kept constant across strata, produced fairly precise relative risk estimates, as reflected by the narrowness of the corresponding confidence intervals  and were assigned one of four exposure levels prior to control sampling, the possibility of some non differential exposure misclassification cannot be ruled out.Neither interviews nor medical data collection were done for the 2170 population controls. We only relied on census-based data, anonymous and parsimonious to protect privacy. We could not therefore verify that controls were "breast cancer free" (but such a misclassification would bias the odds ratio towards the null). Adjustment for age was imprecise, due to the width (20 years) of the age categories used for the control sampling. The lack of information pertaining to residence history and time-activity patterns limited our ability to ascertain the duration of exposure. Considering the long exposure-to-effect interval, in- and out-migrations may have occurred, inducing a potential misclassification. However, as migration is likely to be random with respect to disease status, it would result in a bias of risk estimates towards the null.There are many known risk factors for breast cancer that we were unable to include in our analyses. In particular, women aged 60 years and above exposed to risk factors specific to postmenopausal women  could bOur descriptive analysis of this population based data provides evidence for two main breast cancer populations, according to age at onset, mixed within breast cancer incidence overall. A shift to earlier age at diagnosis resulting from an increased compliance with screening mammography recommendations is unlikely, since the first breast cancer screening campaign in this area started late 2003.One breast cancer type was mostly early-onset with peak incidence near age 50 years. The second breast cancer type was largely late-onset with its mode occurring at age about 70 years. This bimodal age incidence pattern has been observed worldwide, suggesting etiologic heterogeneity due to at least two biological subtypes or causal pathways . In geneOne striking feature lies in the absence of dose-response relationship but the presence of a possible threshold effect at the highest dioxin concentration category (associated with a decreased breast cancer risk). This threshold effect was also noticeable for NHL, but with an increased risk . UnfortuOur results stand in contrast to the findings from two previous studies. A mortality study conducted in Russia reported a higher overall risk of breast cancer  among women living in Chapaevsk . From 19Serum analysis of a subgroup of 981 women living in the high exposure zones around Seveso is also suggestive of an association between TCDD exposure and breast cancer risk . Serum sOur study, on the other hand, is consistent with two different hospital-based case-control studies, that found no association between adipose levels of PCDD/Fs and breast cancer risk ,23. TheyBefore speculating that dioxin anti-estrogenic activity has greater effect on late-onset acquired breast cancer, we must envisage some residual confounding. Considering the inconclusive evidence from studies undertaken so far, future large-scale population based studies that include assessment of family history, breast cancer risk factors, environmental exposures, standardized histopathology reviews and molecular characterization are needed to lead to new insights into the association between environmental dioxin exposure and breast cancer risk.p-dioxins; PCDF: polychlorinated dibenzofurans; TCDD: 2,3,7,8-tetrachlorodibenzo-p-dioxin.Duct NOS: infiltrating duct carcinoma of no special type; GIS: geographic information system; ICD-O: International Classification of Diseases for Oncology; I-TEQ: international toxic equivalency factor; MSWI: municipal solid waste incinerator; PCDD: polychlorinated dibenzo-The author(s) declare that they have no competing interests.JFV conceived and designed the study, and drafted the manuscript. MCC participated in the study design of the study, and carried out the statistical analysis. MH performed the geocoding, and participated in the geographical analysis. SG participated in the geographical analysis. BC carried out the control sampling, and helped interpret the findings. AD participated in the study design, and helped interpret the findings.


Only about 30% of cases of breast cancer can be explained by accepted risk factors. Occupational studies have shown associations between the incidence of breast cancer and exposure to contaminants that are found in ambient air.We sought to determine whether the incidence of postmenopausal breast cancer is associated with exposure to urban air pollution.2) were measured across Montreal in 2005–2006. We developed a land-use regression model to predict concentrations of NO2 across Montreal for 2006, and developed two methods to extrapolate the estimates to 1985 and 1996. We linked these estimates to addresses of residences of subjects at time of interview. We used unconditional logistic regression to adjust for accepted and suspected risk factors and occupational exposures.We used data from a case–control study conducted in Montreal, Quebec, in 1996–1997. Cases were 383 women with incident invasive breast cancer, and controls were 416 women with other incident, malignant cancers, excluding those potentially associated with selected occupational exposures. Concentrations of nitrogen dioxide . Although the size of effect varied somewhat across periods, we found an increased risk of approximately 25% for every increase of 5 ppb in exposure.For each increase of 5 ppb NO2. Further studies are needed to confirm whether NO2 or other components of traffic-related pollution are indeed associated with increased risks.We found evidence of an association between the incidence of postmenopausal breast cancer and exposure to ambient concentrations of NO Breast cancer has the highest incidence rate of all cancers in women and is the second leading cause of death from cancer in both Canada  and the Local vehicular traffic is the primary contributor to air pollution in urban areas. Vehicular emissions include gases, particles, volatile organic compounds, and polycyclic aromatic hydrocarbons (PAHs), many of which are accepted or potential carcinogens. Benzene, for example, is present in gasoline, is an accepted human carcinogen , and has3) were associated with an increased risk of developing postmenopausal breast cancer  compared with exposure to relatively low concentrations . In a second analysis from that study  for postmenopausal women exposed to higher concentrations at the time of first giving birth compared with those exposed at lower concentrations. They found no evidence that exposures at other periods were associated with increased risk. Furthermore, in analyses stratified by smoking status, statistically significant associations were limited to lifetime nonsmokers.Several studies have shown associations between the incidence of breast cancer and occupational exposure to benzene and to PAHs . Given tat study , estimat2), a marker for traffic-related pollution.The purpose of the present study was to determine whether the incidence of postmenopausal breast cancer in Montreal, Quebec, was associated with exposure to intraurban concentrations of nitrogen dioxide  at time of diagnosis. Proxy respondents, who were close family members, completed a total of 75 of the questionnaires. Ethics committees at all participating hospitals and affiliated universities approved the protocol, and signed informed consent was obtained from participating subjects.Occupational exposures were estimated using a standard methodology . BrieflyNeighborhood deprivation may be a confounding factor in the association between breast cancer and air pollution because deprived populations often live in areas that are characterized by higher concentrations of air pollution . Thus, c2 was conducted in 2005 and 2006  were linked to the exposure surface. Ninety-eight percent of subjects were geocoded to the x- and y-coordinates of their home address, and 2% were geocoded to the centroid of the area represented by the six-character postal code, which in Montreal refers usually to a block face or to a large apartment complex.We used these observations to develop a land-use regression model to predict concentrations of mean annual NOe island . We mode2 from Environment Canada’s National Air Pollution Surveillance network .To account for the possibility that the spatial patterns have changed over time, we developed two separate but related methods , 2010 to2 collected at these stations to adjust our land-use regression surface to reflect the spatial patterns of the past. Because of incomplete information, however, observations of annual mean concentrations of NO2 were available at only nine stations for both 2006 and 1985 and at only 10 stations for both 2006 and 1996. We used inverse distance weighting on the mean annual concentrations at these stations, in each year, to interpolate spatial surfaces. Two approaches were used. First, we divided the interpolated surface describing concentrations of NO2 in 1985 by the interpolated surface for 2006. We then multiplied our original land-use regression model by this ratio to produce an extrapolated surface of estimated annual mean concentrations of NO2 for 1985. This process was repeated with the data for 1996. Second, we used the predicted values from our land-use regression surface in 2006 at the locations of the stations for which there were observations in 1985 (n = 10) and in 1996 (n = 12), respectively, and used these data to create new interpolated surfaces. Similar to the first method, we divided the new interpolated surface of the observed concentrations of NO2 in 1985 by the interpolated surface that made use of predicted values (from the land-use regression of 2006), which was then multiplied by the land-use regression surface of 2006. Again, the process was repeated with the data for 1996.The network in Montreal included 13 fixed-site stations that were used to measure hourly concentrations of several criteria pollutants. Our goal was to use the spatial patterns of concentrations of NO2 for 2006 but uses fewer data points, whereas the second is based on interpolated surfaces created with predicted estimates of NO2 for 2006 but is based on a greater number of data points.The key differences between these methods of adjustment is that the first is based on interpolated surfaces created with observed measurements of NO2 during the 10-year period between 1985 and 1996 . These final surfaces were created by adding each 1996 surface to the corresponding 1985 surface and dividing the outcome by 2. To assess the spatial variability between all seven exposure surfaces , 1,000 randomly generated points were sampled and Pearson correlation coefficients were estimated.We also created two additional surfaces that described the mean estimates of NOWe used unconditional logistic regression to estimate ORs and associated 95% CIs. In order not to lose subjects because of missing values for continuous nonoccupational covariates, we used parametric smoothers  in age-adjusted logistic models to view the fitted response functions. Based on the fitted plots, we found cut points for each covariable that defined categories such that the OR within each category was approximately constant. Subjects with missing values were assigned to a “missing information” category. Only age at diagnosis and the two neighborhood ecologic covariates were treated as continuous variables.Fully adjusted models included accepted and suspected risk factors for postmenopausal breast cancer : age at 2 as a continuous, linear variable after verifying this assumption through the use of natural cubic spline functions (2–3 degrees of freedom) and visual inspection of the fitted exposure–response curves. ORs are presented for each increase in exposure to NO2 of 5 ppb (referred to as OR5ppb) and for an increase across the interquartile range for each exposure period . The sensitivity of the results was assessed by analyses limited to those subjects for whom we had information on duration of residence and who had been residents at the same address for ≥ 10 years before diagnosis, by excluding proxy respondents, and by excluding controls with bladder cancer, because there is evidence that exposure to diesel exhaust may increase the risk of incidence of bladder cancer , and the lowest in 2006 (11.3 ppb)  and among observed concentrations of annual mean concentrations of NO2 at the fixed-site stations for 1985, 1996, and 2005 , suggesting that the spatial patterns of NO2 did not vary importantly during the 20-year period between 1985 and 2006. Furthermore, the observed concentrations at the locations of the fixed-site monitoring stations in 2006 were correlated with those in 1996  and in 1985 .Concentrations of NO1.3 ppb) . The treA total of 1,631 subjects were potentially eligible for this study. Interviews were conducted among 608 cases and 667 control subjects, for response rates of 81.1% for cases and 75.7% for controls. Of these 1,275 participants, 106 were deemed to be premenopausal and were therefore excluded; a further 79 participants had incomplete or inaccurate address information, and 291 subjects resided outside of the Island of Montreal. Therefore, these analyses included 799 subjects: 383 cases and 416 controls .The most frequent sites of cancer in the 416 controls were colon (21.6%), uterus (19.0%), ovaries (9.1%), rectum (6.7%), and bladder (6.0%). We found associations for the generally accepted risk factors for postmenopausal breast cancer  . We foun5ppb ranged from a low of 1.05  using estimates of exposure for 1985 to a high of 1.15  in 2006 . The ORs calculated per interquartile range were less variable between exposure periods compared with those computed per 5 ppb, varying from 1.19 in 1985 to 1.30 in 1996 . The two methods of historical extrapolation produced almost identical associations with the risk of developing invasive breast cancer . None of the other covariates included in the fully adjusted models contributed > 3% to the increase in the ORs.Adjustment for individual-level educational status had the strongest effect on the size of the ORs. For example, in a reduced model using estimates of exposure for 2006, adjusted only for age and individual-level educational status, the OR5ppb ranged from a low of 1.23  using estimates for 1985 to a high of 1.52  using estimates for 2006. Similar to the findings with the full data set, the ORs calculated per interquartile range were much less variable between exposure periods, varying from 1.31  to 1.35  .We also computed ORs for the 408 women  who reported that they had been residents at the same address for at least 10 years before diagnosis. These ORs were slightly larger than those produced with the full study population, although because of the smaller sample size, the CIs were wider . For exaWe also computed ORs for subsets that excluded proxy respondents, whose interviews could have elicited less accurate information , and controls diagnosed with bladder cancer, a cancer linked to diesel exhaust fumes exposure . In both analyses, we found associations very similar to those found with the full study population .2 and the incidence of postmenopausal breast cancer. Although the size of effect varied somewhat using estimates of exposure from different periods, we found an increased risk of approximately 25% for every increase of 5 ppb in exposure.We found evidence of an association between exposure to outdoor concentrations of NO2 using independent data sources. The cancer sites in the control group are not known or suspected to be associated with air pollution, with the possible exception of bladder cancer. However, should any of the control series cancer sites be shown subsequently to be causally associated with air pollution, this would have the effect of attenuating the risks demonstrated in our study. The exclusion of bladder cancer, which has been associated with diesel exhaust fumes, did not substantially change the results. Additionally, we found that individual-level educational attainment was the only covariate to affect the estimate of risk by ≥ 5%.Several methodological strengths of our study suggest that the effect sizes are unbiased and perhaps even conservative. Response rates among cases and controls were relatively high, and there is no reason to believe that catchment areas of different hospitals would differ for cases and controls by level of exposure to traffic-related pollution, thus limiting the likelihood of selection bias. The selection of cancer patients as controls reduces the possibility of recall bias for the covariates, and we estimated concentrations of NOWe acknowledge that the present analysis provides only partial information on personal exposure to air pollution. Total personal exposure relates to a number of factors, including daily activity patterns and amount of time spent indoors and outdoors, among others. Two limitations of using the home address as a surrogate of exposure are related to population mobility: People do not necessarily live in the same home over the course of their lifetime, nor do they spend all of their time at home. Although it is true that many subjects may spend their days away from home, a study by Last, given the inherent imprecision associated with geocoded addresses and other geographic data, as well as the fact that we used spatially derived exposures as surrogates for personal exposures, the risk estimates presented here are likely subject to nondifferential misclassification bias. Our results thus probably underestimate the true estimates of the relative risk of postmenopausal breast cancer associated with exposure to air pollution in this population.2 is a causal factor; it is more likely a marker of the complex mixture that is derived from combustion (Our findings should not be interpreted as meaning that NOmbustion .We found an association between exposures to traffic-related air pollution and the incidence of postmenopausal breast cancer in a city that by international standards is relatively unpolluted. Our findings are qualitatively similar to those reported in the two other studies that examined the hypothesis that breast cancer may be associated with exposure to air pollution . Our res


The authors investigated the association between socioeconomic position and stage of breast cancer at the time of diagnosis in a nationwide Danish study. All 28 765 women with a primary invasive breast cancer diagnosed between 1983 and 1999 were identified in a nationwide clinical database and information on socioeconomic variables was obtained from Statistics Denmark. The risk of being diagnosed with a high-risk breast cancer, that is size >20 mm, lymph-node positive, ductal histology/high histologic grade and hormone receptor negative, was analysed by multivariate logistic regression. The adjusted odds ratio (OR) for high-risk breast cancer was reduced with longer education with a 12% reduced risk , 0.80,0.96) in women with higher education and increased with reduced disposable income . There was an urban–rural gradient, with higher risk among rural women  and lower risk among women in the capital suburbs  and capital area . These factors were significant only for postmenopausal women, although similar patterns were observed among the premenopausal women, suggesting a subgroup of aggressive premenopausal breast cancers less influenced by socioeconomic factors. Although affluent women have a higher incidence of breast cancer than socially deprived women, several studies using individual or area-based socioeconomic measures have shown that deprived women with breast cancer have poorer survival from disease  with a primary invasive breast cancer diagnosed between 1 January 1983 and 31 December 1999 and who were less than 70 years of age at the time of diagnosis. The DBCG has since 1977 registered breast cancer patients in Denmark and conducted protocol-based randomised trials of surgery, radiation, chemotherapy or endocrine therapy in patients with primary invasive breast cancer . The regN=460) was considered as unknown in this study.All identified women with breast cancer were classified into a low-risk or high-risk group. The criteria for being a low-risk breast cancer patient has changed over the period and so we redefined a set of criteria consistent with the latest risk protocol to classify all women regardless of the protocol under which they were originally diagnosed and treated. Low-risk breast cancers were defined by tumour ⩽20 mm, no tumour positive axillary lymph nodes, grade of malignancy I or unknown or nonductal tumour, and receptor positive or unknown. Tumour size only reported as <50 mm (Information on socioeconomic characteristics of women with breast cancer was obtained by data linkage to the population based Integrated Database for Labor Market Research (IDA) administered by Statistics Denmark since 1980 . The cor0.6) and deflated according to the 2000 value of the Danish crown (DKK) was categorized as <100 000 DKK/year, 100 000–129 999 DKK/year, 130 000–165 000 DKK/year, >165 000 DKK/year; housing tenure as owner-occupied or rental; and size of dwelling as 0–99 m2, 100–124 m2, 125–149 m2, ⩾150 m2. Demographic variables included age entered as a linear variable, cohabitation status as: single or living with partner, children living at home as: none, 1, 2–5 and degree of urbanization as capital area, capital suburban area, provincial cities and rural areas.Highest attained education was categorized as basic school/high school, vocational training, higher education and unknown; job position as higher functionaries/self-employed, lower functionaries, skilled workers, unskilled workers, not in the work force  and pensioners (retirement and disability); disposable income adjusted for number of people in household ((household disposable income/number of people in household)In Denmark, organized screening programs were initiated in Copenhagen municipality, Funen County, and Frederiksberg municipality in 1991, 1993 and 1994, respectively. Altogether, 100 000 women aged 50–69 years are covered in the three programs, equivalent to some 20% of the total Danish female population in that age group . Since 1977, the NPR has retained key information on all hospitalizations in Denmark, including the personal identification number of the patient, the date of discharge and up to 20 diagnoses and surgical procedures performed during the hospital stay . Tests for interaction (effect modification) between covariates were performed for one pair of covariates at a time using the Wald test statistic. No significant interactions between any pair of variables were observed. In order to explore the influence of menopausal status on the association between socioeconomic position and high-risk breast cancer, we stratified the cohort according to menopausal status and ran the logistic regression analyses separately in premenopausal and postmenopausal women.Logistic regression models were developed to examine the simultaneous influence of all socioeconomic and demographic factors of interest on the likelihood of being diagnosed with a high-risk breast cancer using the procedure PROC LOGISTIC in SAS 9.1. on a UNIX platform. Age was entered linearly in the logistic regression model. This is biologically more reasonable than the step functions corresponding to categorization and furthermore, increases the power of the analysis . Our expN=11 685) or postmenopausal (N=17 080) with six women of unknown menopausal status below age 55 classified as premenopausal and one over age 55 as postmenopausal.By use of the risk group definition, 6007 women were diagnosed with low-risk breast cancers and 23 808 women with high-risk breast cancers. Some 1955 women could not be classified due to unknown tumour size (971 women or 3%) or number of positive lymph nodes  and were thus excluded from the analyses. Some 360 women (1%) diagnosed with any tumour prior to the breast cancer according to the files of the NPR were excluded in order to be consistent with the criteria of no previous malignant disease in DBCG. A further 702 women (2%) were excluded due to missing information in one or more of the socioeconomic variables leaving a total of 28 765 women for analyses. All women were categorized as either premenopausal (vs 16%), who had basic school or high school education (48 vs 44%) or low disposable income (27 vs 24%) were diagnosed with high-risk breast cancer. Further, more women with 2 or more children living at home (10 vs 7%), lived in rural areas (35 vs 30%) and with access to screening (93 vs 88%) were diagnosed with high-risk breast cancer.Characteristics of the 28 765 women included in the study population with 5809 (20%) women classified as having low-risk breast cancers and 22 956 (80%) women classified as having high-risk breast cancers are shown in The multivariate analysis showed that the odds ratio (OR) for being diagnosed with a high-risk breast cancer was reduced by 5% per year up to age 44 years  after which age the risk was constant . FurtherIn premenopausal women the effect of age on risk of being diagnosed with a high-risk breast cancer was similar to that found in the full analysis, with a 5% reduction per year up until age 44 years after which age the risk was close to constant in both premenopausal and postmenopausal women . As therThis population-based Danish study shows an increased risk for being diagnosed with a high-risk breast cancer with shorter education, with lower disposable income, with a residence in rural areas, and with having no access to organized mammography screening. Apart from the access to screening, these effects of social inequality were significant only for postmenopausal breast cancers.The strengths of this study include the unselected cohort of breast cancer patients covering the entire Danish nation and the possibility to define disease progression at the time of diagnosis by four clinical criteria based on information from a nationwide clinical database on breast cancer in Denmark. Further, all information on the socioeconomic and other variables has been collected prospectively and uniformly for administrative purposes independently of our study hypotheses, thus eliminating recall bias and information bias. The tradition for administrative registration in Denmark further enabled us to identify partners and children living at home and thereby adjust disposable income by number of persons in the household.As expected, we find that access to mammography screening is an important factor for breast cancer stage at diagnosis. Mammography screening is not available throughout the country and furthermore, attendance rates have been reported to average 66% in Copenhagen and 84% in Funen through screening rounds 1–3  (We measured socioeconomic position by education, occupation, income and housing acknowledging that each of these indicators measured different although often related aspects. Although the education level might have undergone considerable changes in the women included in our study with more older women in the early study period attaining only basic school education, the similar trends in both pre- and postmenopausal women by education indicate that the categorisation captures an effect probably related to knowledge and skills that in turn affects cognitive functioning (The social inequality at diagnosis observed in our study contrasts with the increasing risk for breast cancer with increasing education . This deWe included marital status (The social inequality in risk group of breast cancer by education and disposable income were mainly in postmenopausal women, although similar patterns of increased progression of tumour at diagnosis with shorter education or lower income were observed among the premenopausal women as well. One interpretation of this might be that the more aggressive breast cancers which are characterized by young age at manifestation and genetic predisposition which are predominantly found among premenopausal women are less influenced by the effect of social deprivation on tumour progression.This population based study shows that less education, disposable income, and degree of urbanization increase the risk of being diagnosed with a high-risk breast cancer in Denmark and that access to mammography screening reduces this risk. Apart from access to mammography screening, these associations seemed stronger in postmenopausal women, suggesting that a subgroup of aggressive premenopausal breast cancers are less influenced by the effect of socioeconomic position on tumour progression at time of diagnosis.