N=139 breast cancer cases), we found no association between smoking before the birth of a first child and breast cancer risk: fully adjusted  odds ratio 1.06 . The pooled estimate from a meta-analysis of our study and 11 previously published studies (N=6528 cases) was 1.07 . We conclude that smoking prior to the birth of a first child is not associated with increased risk of breast cancer.It has been suggested that the period between puberty and first birth is a time when the breast is particularly susceptible to carcinogenic effects. In a cohort of 3047 women aged 60–79 years ( It has been suggested that smoking may play differing roles in the aetiology of breast cancer depending upon the period in a woman's life during which she smokes . Breast A small number of studies that have assessed the association of smoking prior to completion of first pregnancy with breast cancer and have reported conflicting results . We aimeData from the baseline assessment and first 3 years of follow-up of the British Women's Heart and Health study were used. Full details of baseline assessment have been reported previously . In brieIn this paper, we have included all cases of breast cancer reported at baseline assessment  and those occurring over the median 3.5 years follow-up period (incident cases). Three main sources were used to determine breast cancer status: (i) self-report in the baseline questionnaire and Q3; (ii) a diagnosis recorded in the medical records and (iii) all participants were flagged with the National Health Service central register (NHSCR) that provided details of cancer registrations. Anyone with a diagnosis of breast cancer from any one of these three sources was considered to be a case. Five additional incident cases that had not been identified by any of these three sources were obtained because they occurred as the underlying cause or elsewhere on the woman's death certificate. Thus, this study is of prevalent and incident breast cancer cases occurring any time in the woman's life up to January 2004.Age at menopause, age at menarche, number of pregnancies and live births, current and past use of hormone replacement therapy (HRT) and past use of oral contraception, history of a hysterectomy and/or oophorectomy, alcohol and smoking history were obtained from the self-completed questionnaire and/or the research nurse interview . DetaileP-values.Multiple logistic regression was used to assess the association between smoking in relation to first birth and breast cancer with adjustment for potential confounding factors: age (entered as a continuous variable), age at first birth , age at menarche , number of births , age at menopause , use of HRT (ever or never), hysterectomy and/or oophotectomy (binary), childhood and adult social class , body mass index (BMI) (continuous) and alcohol consumption . To further explore whether smoking around the time of breast development was a risk factor for future breast cancer, we assessed the association of smoking during the period 1 year before menarche and the first 5 years after menarche in all women using multiple logistic regression. For all analyses robust standard errors, which take into account possible nonindependence between women from the same town, were used to calculate confidence intervals (CI) and a priori to pool estimates using DerSimonian and Laird's random effects methods, since it was likely that there would be heterogeneity between studies due to differences in the number of pre- and postmenopausal cases and differences in study design (Searches of Medline and Embase (up to January 2004) were undertaken using extended terms for breast neoplasia and smoking. Any published study that assessed the association between smoking before a first pregnancy and breast cancer risk was included in a meta-analysis. We decided y design . MetaregOf the 4286 women in this study, 225 (5.3%) had breast cancer ascertained from at least one source. The majority (82%) of these cases were identified from at least two sources. The age distribution of women with cancer identified by each source were similar – mean (standard deviation) age of women with breast cancer identified by self report 68.7 (5.3), identified by medical record review 68.9 (5.3) and identified by cancer register 69.0 (5.4). Of the 225 cases, 170 were prevalent and 55 incident cases.Of the 4286 women, 3467 (81%) had at least one live birth and of these 3047 (88%) gave their age at the time that their first child was born. The associations presented in Among all women in the cohort (4286 women with 225 cases of breast cancer), the age-adjusted odds ratio for smoking either in the year prior to or within 5 years of menarche compared to never smoking was 0.96. ; the fully adjusted odds ratio was 1.00 .In the whole cohort, 186 (83%) of the total 225 cases were postmenopausal and in the subgroup of women with at least one birth and a known age at first birth 118 (85%) of the 139 breast cancers were postmenopausal. We were therefore unable to determine with any level of precision the effect of associations on premenopausal cancers. The associations for postmenopausal breast cancers only did not differ from those presented and the point estimates for premenopausal cancers were similar to those for postmenopausal cancers and close to the null value . When the analyses were restricted to incident cases only, although considerably less precise, the results did not differ from those presented: fully adjusted odds ratio 1.08 . When all analyses were repeated using breast cancer data from each of just one of the three sources , the results were unchanged.P=0.01), which was not explained by menopausal status (P=0.34). The Egger test did not suggest strong evidence of small study bias (P=0.23), although the Begg test provided more evidence of this (P=0.07). The funnel plot , the pooled odds ratio (95% CI) was 1.03  with no strong evidence of heterogeneity between studies (P=0.23). Other than being two of the smallest and least precise studies, it can be seen from We identified 11 studies in 10 publications; these are summarised in Our results suggest that there is no association between a woman smoking before the birth of her first child and breast cancer. Our finding that smoking around the time of puberty was not associated with increased risk of breast cancer is consistent with a number of other studies. the results were unchanged. By combining information from all three sources, it is likely that most cases have been identified.An important limitation of our, and other, studies is the small number of cases and hence imprecision of the results. This highlights the importance of pooling results from all studies in order to provide a precise estimate of the overall effect.a priori hypothesis, our results suggest that smoking solely after a woman's first birth may be protective against breast cancer. Two other studies have also found reduced risk of breast cancer associated with smoking only in the period after a first birth (Although imprecise and not an st birth . Some cost birth , and twoSmoking around the time of breast tissue development and before full differentiation – between puberty and prior to completion of the first pregnancy – is not associated with increased risk of breast cancer. Smoking is associated with a number of adverse health outcomes and young women should be discouraged from taking up smoking irrespective of any association or lack of association with breast cancer.


Sir,We found Dr van Noord's preconceptional mitochondrial hypothesis interesting particularly in line with a recent report linking polymorphisms of two DNA base excision repair genes (XRCC1 and hOGG1) to breast cancer risk in daughters born to older mothers . MnSOD hTo test this hypothesis, we analysed the association of parental age with breast cancer risk using data from the Shanghai Breast Cancer Study (SBCS), and the results are in shown Although we collected information on whether the mother had a threatened miscarriage with the index pregnancy, too few women reported this adverse event  to provide a stable risk estimate. Other studies may have sufficient numbers of mothers to investigate this aspect of Dr van Noord's hypothesis.Dr van Noord argued that insulin-like growth factor- I (IGF-1) might be unlikely to explain the inconsistent findings on birth weight and breast cancer risk in the literature, since the link between IGF-I and breast cancer risk has been found primarily in premenopausal women, while the high birth weight-breast cancer association has been seen among pre- and postmenopausal women. A previous report from the SBCS showed that elevated levels of IGF-I were associated with an increased risk of breast cancer among all women, but the association was more pronounced among women diagnosed premenopausally and among women with a high body mass index or waist-to-hip ratio (


We have previously found an increased risk of breast cancer among women born preterm. To confirm or refute the results, an enlarged study was conducted. The results from this study do not confirm the initial findings and suggest that preterm birth can be ruled out as a risk factor for breast cancer. Oestrogens are fundamental hormones in breast cancer aetiology, and it has been suggested that the surge in oestrogens during pregnancy may influence subsequent breast cancer risk in the offspring .Infants born extremely preterm, that is before 28 weeks of gestation, has been found to have high levels of oestrogens postnatally . Approximately 250 000 birth records were examined. Female subjects born before 35 completed weeks of gestation according to date of last menstrual period, and females born after 35 weeks with a birth weight of less than 2000 g were included. At the point of initial inclusion into the cohort, data on pregnancy and delivery, such as date of last menstrual period and birth weight, was recorded.By use of the National Registration Number, the cohort was followed up in the Swedish Register of Population and Population Changes from 1 January 1958 to 31 December 1998 or to death or emigration prior to this time. Subjects who deceased before 1958 were excluded, since incident cancers could not be identified with any completeness prior to the initiation of the Swedish Cancer Register. We identified cases of cancer by linking our subjects to the Swedish Cancer Register. The completeness of this register is over 98% .n=155), and subjects with a gestational duration of 33–34 weeks with a birth weight of 3600 g or more (n=94).We used last menstrual period to estimate gestational duration. We categorised gestational duration into three groups: 32 completed weeks or less, 33–34 weeks, and 35 weeks or more. Birth weight was categorised into less than 2000 g, 2000–2999 g, and 3000 g or more. Since an unreasonably high birth weight for a short duration of gestation indicated an underestimation of gestational duration or an overestimation of birth weight, we excluded subjects born before 32 weeks of gestation with a birth weight of 3000 g or more  for all female cancer, female breast cancer, and early onset female breast cancer using age, sex, and time-specific cancer incidence rates for the Swedish population during the period 1958–1998 provided by the Cancer Registry. We analysed all cancer occurrences as well as the first cancer occurrence. As these estimates were essentially the same, we only present the results from the analysis of first cancer occurrence.There were 1483 women who met the eligibility criteria and of these, 1208 were born before 35 weeks of gestation and 275 were born after week 35 with a birth weight of less than 2000 g (The overall risk of cancer among the women was not increased (This study is, to our knowledge, the largest cohort study hitherto conducted on long-term cancer risk among women born preterm or with low birth weight. Neither preterm birth nor low birth weight were associated with an increased risk for cancer overall or for breast cancer.We have previously reported an increased risk for breast cancer among women born before 32 weeks of gestation(Consistent with other studies (When The use of self-reported date of last menstrual period leads to some misclassification in the calculation of gestational duration. The nature of this misclassification is both nondifferential and systematic, the latter with a tendency to overestimate gestational duration(In conclusion, we found that neither preterm birth, nor low birth weight, are associated with an increased risk for breast cancer.


The International Agency for Research on Cancer classifies acrylamide as a probable human carcinogen. It has only been in recent months that an epidemiologic study first found a link between dietary acrylamide and human cancer risk. Now Danish researchers report that acrylamide adduct levels in blood are associated with an increased risk of breast cancer in postmenopausal women.This is the first epidemiologic study to use blood biomarkers to assess acrylamide exposure. The findings, says first author Pelle Thonning Olesen, emphasize the importance of using biomarkers for exposure assessment. “Biomarkers are a more trustworthy indicator for exposure,” he says.Before 2002, people were known to be exposed to acrylamide in certain industries and through smoking tobacco. That year, the Swedish National Food Administration discovered that acrylamide also forms in fried or baked starchy foods such as french fries, coffee, and baked goods. Diet is now thought to be the major source of exposure among nonsmokers, but the cancer risk posed by acrylamide in food is unknown.All previous epidemiologic trials estimated acrylamide consumption from food frequency questionnaires. Olesen, a toxicologist at the Technical University of Denmark, Søborg, and colleagues instead measured levels of acrylamide and a key metabolite, glycidamide, bound to hemoglobin. The subjects included 374 postmenopausal women with breast cancer and 374 controls who participated in the Danish Cancer Society’s Diet, Cancer, and Health Study.Adduct levels of acrylamide among smokers reflect both dietary and smoking intake of the compound. After statistical adjustments for smoking behavior, women with the highest acrylamide-hemoglobin levels showed a 2.7 times higher risk of estrogen receptor–positive breast cancer compared with women with the lowest acrylamide-hemoglobin levels. The risk rose with increasing acrylamide exposure.International Journal of Cancer.Acrylamide-hemoglobin levels were not linked to estrogen receptor–negative breast cancer, and glycidamide-hemoglobin levels showed no connection with any breast cancer. The finding that only acrylamide-hemoglobin was associated with breast cancer suggests that the compound may induce cancer by nongenotoxic routes such as alkylation of proteins that could alter estrogen receptor function. The findings were reported in the 1 May 2008 issue of the “This is an important study because it’s the first to measure acrylamide adducts,” says epidemiologist Lorelei Mucci, an assistant professor at the Harvard School of Public Health. Nonetheless, the study should be repeated in larger numbers of nonsmoking women, according to Mucci, because more than half the cases and controls were current or former smokers.“The big public health question here is whether the amount of acrylamide in foods is enough to lead to cancer,” Mucci says. It is possible that other chemical compounds formed along with acrylamide may be the culprit in any cancer link. “Acrylamide-hemoglobin may be a biomarker for other carcinogenic chemicals formed during the heating of foods,” cautions Olesen.


CYP19 Arg264Cys and CYP1B1 Leu432Val polymorphisms on breast cancer risk in a series of Korean breast cancer patients and controls. The results suggest that the CYP19 Arg264Cys polymorphism modifies breast cancer risk , especially in association with alcohol consumption (P for interaction=0.04), whereas the CYP1B1 Leu432Val polymorphism appears to play no role here.A case–control study was performed to assess the potential influence of  Breast cancer is the second most frequent cancer in Korean women and its incidence is increasing . A varieCYP19 gene encodes aromatase, which catalyses the formation of oestrogens from testosterone and androstenedione. To date, several polymorphisms have been found in the CYP19 gene  was collected using a questionnaire administered by trained interviewers.The criteria of subject selection and details of data collection on lifestyle have been described elsewhere . Briefly, a 63 bp long amplification product was obtained using about 10 ng DNA as template in a polymerase chain reaction (PCR) with 20 pmol of oligonucleotide primer P1 (5′-TGC CAT AGA AGT TCT GAT AGC-3′) and 4 pmol of primer P2 (biotin-labelled 5′-TCT TCC AGT TTC TCT TCT GT-3′)  in a total volume of 20 μl. The amplification conditions were: initial denaturation at 95°C for 5 min followed by 45 cycles of 15 s at 94°C and 30 s at 58°C . The PCR products were hybridised with the following probes: 5′-AAA AAA GAC GCA GGA TT-3′ and 5′-AAA AAA GAT GCA GGA TT-3′. The CYP1B1 was genotyped for 241 cases and 290 controls recruited from 1995 to 1999 by a real-time PCR-based method as described by DNA was isolated using standard methods from blood drawn into 10 ml heparinised tubes and stored at −70°C until use. The χ2 on one degree of freedom.Odds ratios (ORs) and 95% confidence intervals (CIs) were estimated by unconditional logistic regression analysis adjusting for age, education, body-mass index, age at first full-term pregnancy, family history of breast cancer, duration of breast feeding, and alcohol consumption. Information on alcohol consumption was collected by three questions: (1) How frequently do you drink alcohol ? (2) How long have you been drinking? (3) Have you ever quitted drinking? Alcohol intake was categorised as: never-drinking (never and less than once per month) or ever drinking (at least once per month). Interactions between known risk factors  and each genotype were evaluated by likelihood ratio test. The difference of two −2log L values of logistic models with and without interaction terms was referred to tables of CYP19 and CYP1B1 genotypes are shown in Table 2CYP19 Cys allele-containing genotypes (39%) was somewhat lower than that previously found in Japanese women (52–54%) , on the other hand, was similar to that (29%) found in Japanese women , whereas the CYP1B1Val allele-containing genotypes had no effect in this context  (The 0.7–1.6) .CYP19 genotype and alcohol consumption (P for interaction=0.044). Ever-drinking women with CYP19Cys allele-containing genotypes showed a 3.3-fold risk (95% CI=1.7–6.5) for breast cancer compared with never-drinking women with the Arg/Arg genotype , a significant interactive effect was observed between CYP19 genotypes, no interaction was observed between the known risk factors for breast cancer and the CYP1B1 genotypes. Neither was there a gene–gene interaction between the CYP19 and CYP1B1 genotypes.In contrast to the CYP19 genotype and alcohol consumption play important roles in breast cancer development, and that these factors could synergistically increase the risk of breast cancer in Korean women. The contrasting findings with previous studies , and different genotyping methods .Our results suggest that both 264Cys polymorphism is located in or near recognition site of CYP19 aromatase and thus, it might enhance the oestrogen synthesis and exposure to endogenous oestrogen. However, it is also be possible that CYP19 Arg264Cys polymorphism is in linkage disequilibrium with other important polymorphic site such as TTTA repeat polymorphism .In contrast to the CYP19 genotype and alcohol consumption play important roles in breast cancer development and that these factors may synergistically increase the risk of this malignancy in Korean women. Epidemiological studies with large sample sizes are, however, required to confirm these preliminary findings and to evaluate the role of the other CYP19 genotypes of potential interest in breast cancer development in this ethnic group.In conclusion, the results of this study support the hypothesis that


BRCA1 and BRCA2 mutations increase breast and ovarian cancer risks substantially enough to warrant risk reduction surgery, despite variable risk estimates. Underlying this variability are methodological issues, and also complex genetic and nongenetic effects. Although many modifying factors are unidentified, known factors can already be incorporated in individualised risk prediction. BRCA1 and BRCA2 were identified as genes mutated in hereditary breast/ovarian cancer by genetic analysis in families with multiple cases of these malignancies. In the following decade, as BRCA1 and BRCA2 testing became more common in patients with a personal or family history of cancer, numerous studies assessed cancer risks in carriers of these mutations. These studies have resulted in variable risk estimates, leading to difficulties in defining risk figures that are clinically applicable to the individual patient. However, they illuminated the complexities that underlie risk prediction even in the context of a single gene with major effects on cancer predisposition. The emerging picture is that breast and ovarian cancer risks in BRCA1 and BRCA2 carriers are substantially higher than in the general population, but that they are considerably affected by nongenetic, environmental factors, and by additional genetic modifiers.vs true carrier status, and from errors in cancer reporting by relatives, a particularly common problem for ovarian cancer. In a meta-analysis of such case-based studies, by age 70 years, in BRCA1 carriers breast cancer risk was 65% (95% CI 51–75%) and ovarian cancer risk was 39% (95% CI 22–51%), and in BRCA2 carriers breast cancer risk was 45% (95% CI 33–54%) and ovarian cancer risk was 11% (95% CI 4.1–18%)  families, by age 70 years, BRCA1 carriers had a breast cancer risk of 71% (95% CI 53–82%) and an ovarian cancer risk of 47–63% (4.1–18%) . These bBRCA1 and BRCA2 are large genes, with thousands of different mutations. A clear genotype–phenotype correlation exists for BRCA2 mutations, where the central part of the gene (nt 3035–6629 in exon 11) is named the ovarian cancer cluster region (OCCR) because mutations within it are associated with an increased ovarian: breast cancer ratio. Compared to other BRCA2 mutations, OCCR mutations are associated with higher ovarian cancer risk  and lower breast cancer risk  . SimilarThe phenotypic expression of inherited disease can vary widely even between individuals with the same disease-causing mutations, as a result of both nongenetic variation and polymorphisms/mutations in other genes, known as genetic modifiers. There is epidemiological evidence for the existence of such modifiers for BRCA1 and BRCA2 mutations, since there is familial clustering of the cancer site: carriers in families with ovarian cancer index cases are at higher risk for ovarian cancer and at lower risk for breast cancer than carriers from families with breast cancer index cases , but thiThere are very significant nongenetic effects on breast and ovarian cancer risk in BRCA1 and BRCA2 carriers. This is evidenced by the consistent observation of increasing breast and ovarian cancer risks in carriers born in more recent years . Risk reYounger age at menarche is associated with increased risk for sporadic breast cancer. An effect was not observed in BRCA2 carriers, but BRCA1 carriers whose age at menarche was 14–15 years had a 54% reduction in breast cancer risk compared to those with menarche at ⩽11 years of age  (vs nulliparous carriers)  breast cancer in BRCA1 carriers , a largeIn a case–control study of 965 BRCA1 and 280 BRCA2 pairs, breast feeding did not influence breast cancer risk in BRCA2 carriers, but BRCA1 carriers who breast fed for over 1 year were less likely to have had breast cancer than those who never breast fed   breast cancer was increased in women who first used OC before 1975 , who used OC before age 30 years , or who used them for 5 or more years  . In a cavs the lowest tertile increased breast cancer risk . A larger case-only study found that increased physical activity and lack of obesity in adolescence were associated with significantly delayed breast cancer onset in BRCA1 and BRCA2 carriers . Risk was increased in carriers aged 40 years and younger , particularly those exposed only before the age of 20 years    is a major characteristic of the BRCA2-associated cancer syndrome. Cumulative risk for this malignancy was assessed in 164 BRCA2-BCLC families, which included 59 cases of male breast cancer, and risk was estimated at 2.8% (95% CI 0.6–13.0%) by age 70 years, rising to 6.9% by age 80 years (95% CI 1.2–38.6%) (2–38.6%) . This co2–38.6%) . While BProstate cancer risk is also higher in BRCA2 compared to BRCA1 carriers. Cumulative prostate cancer risk assessed in 173 BCLC-BRCA2 families was 7.5% (95% CI 5.7–9.3%) by age 70 years, corresponding to a RR of 4.65 (95% CI 3.48–6.22), which may include a detection bias. A Dutch study of BRCA2 carriers reached similar conclusions, with a cumulative risk of 5.2% (95% CI 1.7–8.7%) by age 70 years, reflecting an RR of 2.5 (95% CI 1.6–3.8) (1.6–3.8) , and mut1.6–3.8) . The low1.6–3.8) . Studiesvs expected number of cases of a specific malignancy, and studies which compare rates of BRCA1/BRCA2 mutations in unselected cases of a specific malignancy with mutation rates in controls or in the general population. In family-based studies, diagnoses other than breast or ovarian cancer were confirmed by pathology or clinical records in only about half of the cases  in men, and 1.26% (95% CI 0.92–1.72%) in women, reflecting a two-fold RR . In BRCABreast and ovarian cancer risk in BRCA1 and BRCA2 carriers is a complex trait, affected by genetic modifiers and nongenetic factors. While further studies are necessary to define and quantify risk-factor effects, even now the best approach for the individual patient is perhaps a personalised one, which takes into account family history, ascertainment, and known environmental risk factors.


An analysis of the occurrence of breast cancer in this long-term prospective cohort study shows a significant relative risk (RR) in women who have ever used oral contraceptives (OC) of 3.33 in women age 30 to 34 years at diagnosis and an RR of 5.88 (P = 0.0011) in women who were parity 1 at the time of diagnosis. In women below the age of 35 years the RR of 2.38 was not significant. There was no increased risk in women over the age of 35 years. A significant trend relating to duration of use was demonstrable in women who were parity 1 in the analysis of both current and ever-users. An analysis by time since stopping OC use revealed a significant trend in all ever-users, but the trends were much steeper in women of parity 1 or aged 30 to 34 years at diagnosis. There was no evidence that the increased rates in OC users were related to the oestrogen or progestogen dose. The 5 year survival rate in users diagnosed under the age of 35 years was significantly poorer than in comparable non-users. It is possible that the increased rates in younger OC users might be due to an accelerated presentation of breast cancer in those women who would otherwise have been diagnosed at a later time. The non-significant excess risk in users under 35 years of age was approximately 1 in 7,000 users per year. The unresolved discrepancies between the results of the published studies make it impossible at the present time to decide whether or not OC use is associated with an increased risk of breast cancer.


Breast self-examination (BSE) is widely recommended for breast cancer prevention. Following recent controversy over the efficacy of mammography, it may be seen as an alternative. We present a meta-analysis of the effect of regular BSE on breast cancer mortality. From a search of the medical literature, 20 observational studies and three clinical trials were identified that reported on breast cancer death rates or rates of advanced breast cancer (a marker of death) according to BSE practice. A lower risk of mortality or advanced breast cancer was only found in studies of women with breast cancer who reported practising BSE before diagnosis . The results are probably due to bias and confounding. There was no difference in death rate in studies on women who detected their cancer during an examination . None of the trials of BSE training (in which most women reported practising it regularly) showed lower mortality in the BSE group . They did show that BSE is associated with considerably more women seeking medical advice and having biopsies. Regular BSE is not an effective method of reducing breast cancer mortality. For many years, women have been taught methods of breast self-examination (BSE) and it is recommended that they practise this regularly . HoweverRecently, the Studies that reported on rates of death from breast cancer or rates of advanced breast cancer (a marker of death) according to BSE practice were identified from Medline, Embase and Cancerlit (1966–2002), and included in the analysis. Keywords used were ‘breast cancer’ with ‘BSE’ or ‘self-examination’. In some studies, women were classified according to whether they practised BSE regularly or not. In other studies, women were classified according to the method of detecting the cancer: during BSE, by chance (e.g. while washing or dressing), mammography or examination by physician. Below we describe the main common features of the studies, but in the interest of brevity we do not provide further details, since these can be obtained directly from the individual published reports.We included results on mortality or, as a surrogate for death, advanced breast cancer . Analyses are presented separately for these two outcomes.The following types of studies were included in the analyses.Women who reported practising BSE or not and were then followed up for several years  to see who later died from breast cancer.Women who reported whether they found their cancer during self-examination or by chance and were then followed up for several years to see who later died from breast cancer.Women found to have advanced breast cancer at the time of initial diagnosis who reported retrospectively on whether they practised BSE or not.Women found to have advanced breast cancer at the time of initial diagnosis who reported retrospectively on whether they had found their cancer during self-examination or by chance (e.g. washing and dressing).A total of 15 studies were based only on women newly diagnosed with breast cancer  84% had attended all three training sessions. The reports from the Russian trial were based on women who had received training in BSE.In the studies based on only women newly diagnosed with breast cancer, the definition of BSE practice varied. It was monthly  or threeThe relative risks (or odds ratio) and 95% confidence intervals (CI) were estimated from the data in each study. They were pooled on a log scale and weighted by the inverse of the variance, with allowance for any heterogeneity .Table 1P<0.001) in those who practise BSE. There was no evidence of heterogeneity between the studies (P=0.41). If the study in which some women had mammo-graphy (P<0.001). In those women who reported that their cancer was detected during self-examination, there was no evidence of a reduction in the risk of death compared to those who found their cancer by chance . Again there was no strong evidence of heterogeneity between the results (P=0.26). If the study (P=0.98).Figure 1P<0.001). Although there was evidence of hetero-geneity (P<0.001), all the studies reported a reduction in risk. In women who found their cancer during an examination, there was a 34% reduction in risk .Figure 2P=0.42). None of the case–control studies found statistically significant effects with only one suggesting a benefit . After 10 years, the two trials (Russia and China) show that overall there were 53% more biopsies in women who were taught BSE compared to those who were not, this was highly statistically significant . The trials also suggest that at 5 years, women taught BSE were no more likely to be diagnosed with breast cancer than those not taught BSE (relative risk 1.01). After a longer follow-up (9–10 years), there is an indication from the Russian trial that more cancers were found in the BSE group (24% more women diagnosed with breast cancer), but this was not found in the trial from China.In the Russian trial, twice as many women in the BSE group sought medical advice compared to the non-BSE group , and thiP=0.79); there was no evidence of heterogeneity (P=0.94). The results were not materially different if the nonrandomised trial from the UK was excluded, pooled relative risk 1.05 .The risk of dying from breast cancer was remarkably consistent between the three trials and over the different follow-up periods. There was no evidence of an advantage in the BSE group after any length of follow-up. The pooled relative risk was 1.01 with narrow 95% confidence limits . Several studies have shown that various characteristics that are likely to be associated with dying from breast cancer were also associated with BSE practice, but analyses adjusting for the potential effect of such confounding on mortality were not reported. Women who practised BSE tended to be younger, premenopausal and of a higher socioeconomic status .The results presented here on BSE may have an impact on the current debate over the use of mammography screening. Despite clear evidence to the contrary ,2001. Br


Historically, the incidence rate of breast cancer among non-Hispanic white women living in the San Francisco Bay area (SFBA) of California has been among the highest in the world. Substantial declines in breast cancer incidence rates have been documented in the United States and elsewhere during recent years. In light of these reports, we examined recent changes in breast cancer incidence and risk factor prevalence among non-Hispanic white women in the SFBA and other regions of California.Annual age-adjusted breast cancer incidence and mortality rates (1988 to 2004) were obtained from the California Cancer Registry and analyzed using Joinpoint regression. Population-based risk factor prevalences were calculated using two data sources: control subjects from four case-control studies (1989 to 1999) and the 2001 and 2003 California Health Interview Surveys.In the SFBA, incidence rates of invasive breast cancer increased 1.3% per year  in 1988–1999 and decreased 3.6% per year  in 1999–2004. In other regions of California, incidence rates of invasive breast cancer increased 0.8% per year  in 1988–2001 and decreased 4.4% per year  in 2001–2004. In both regions, recent (2000–2001 to 2003–2004) decreases in invasive breast cancer occurred only in women 40 years old or older and in women with all histologic subtypes and tumor sizes, hormone receptor-defined types, and all stages except distant disease. Mortality rates declined 2.2% per year  from 1988 to 2004 in the SFBA and the rest of California. Use of estrogen-progestin hormone therapy decreased significantly from 2001 to 2003 in both regions. In 2003–2004, invasive breast cancer incidence remained higher (4.2%) in the SFBA than in the rest of California, consistent with the higher distributions of many established risk factors, including advanced education, nulliparity, late age at first birth, and alcohol consumption.Ongoing surveillance of breast cancer occurrence patterns in this high-risk population informs breast cancer etiology through comparison of trends with lower-risk populations and by highlighting the importance of examining how broad migration patterns influence the geographic distribution of risk factors. A striking feature of breast cancer epidemiology is its geographic variation in occurrence, with differences in invasive breast cancer incidence as high as 10-fold internationally . At the Prior studies have suggested that most ,7, if noWith the recent and widely publicized changes in breast cancer incidence rates in this region  from the Surveillance, Epidemiology, and End Results (SEER) program of the National Cancer Institute (NCI) for the period 1 January 1973 to 31 December 2003 – the most recent year for which cancer data were complete at the time of this analysis. Analyses were based on incident cases of breast cancer  the California Health Interview Surveys (CHISs) conducted in 2001 and 2003 (source: 2001 and 2003 CHISs) and (b) controls identified through random digit dialing in four population-based case-control studies conducted between 1990 and 2000 in the SFBA . We restricted analyses to the age group (35 to 74 years) common to the four population-based case-control studies: (a) Marin County Breast Cancer Study of Adolescent Risk Factors, which collected information up to 1 year prior to the selection dates, 1997 to 1999, from 297 controls who were residents of Marin County and who did not have a history of breast cancer ; (b) the2) stratified by age group .Breast cancer risk factors included in this study have been associated with breast cancer in prior studies  and eithp values of less than 0.05 considered significantly different. APCs across the period 1988 to 2004 were calculated by fitting a least squares regression line to the natural logarithm of the rates as the outcome variable, with calendar year as the predictor variable [SEER*Stat software  was usedvariable . Time trvariable .AskCHIS application. CHIS data are weighted to the California Department of Finance estimates of the number of residents in each California county by age, race, and gender and the 2000 Census of Population counts from the US Census Bureau [Prevalence estimates from the population-based controls were age-adjusted to the 2000 US standard million population using SAS version 9 software . Age-adjusted prevalence estimates from the 2001 and 2003 CHISs were obtained from the internet-based s Bureau . When thin situ and invasive breast cancer were higher in non-Hispanic white women in the SFBA than in non-Hispanic white women in the rest of California , incidence rates of invasive breast cancer in the SFBA have been consistently higher than rates in other regions since the inception of continuous cancer surveillance by the SEER program in 1973 , but not tumors less than 2 cm in diameter . In 2003–2004, incidence rates of breast cancer in situ and invasive breast cancer remained significantly higher in the SFBA than in the rest of California by 12.5% and 4.2%. From 2000–2001 to 2003–2004, rates of breast cancer 2 (women under 50 years of age), to have been physically active in the past 30 days, to have consumed alcohol in the past month (particularly two or more drinks per day), and to have had their first child after the age of 30 years or to be nulliparous , suggest that the SFBA excess could be due in part to higher levels of screening, a finding supported by CHIS data that find a somewhat higher prevalence of mammographic screening in 2001, but not 2003, in the SFBA. The continued assessment of future trends in incidence rates will help us to understand whether a plateau in mammography screening [It is unclear to what extent mammographic screening patterns, which have been associated with breast cancer incidence increases in the US  level rather than at the individual level. In addition, we did not adjust breast cancer incidence rates for known risk factors, nor did we have population prevalence estimates of risk factor changes over the 17-year study period or in the rest of the SEER regions. We were able to present risk factor changes between variables measured similarly in the 2001 and 2003 CHISs, but these data are limited by low response rates that could result in selection bias. However, response rates in CHIS were comparable to those in other population-based surveys , and resUnderstanding breast cancer incidence and risk factor prevalence patterns in SFBA non-Hispanic white women informs the study of breast cancer etiology in two ways. First, future population-based studies attentive to exposure heterogeneity might include SFBA counties together with other US or international populations with lower documented incidence rates of breast cancer, in which established or putative risk factors may be less common, and protective factors may be more common. Second, high-incidence populations are also useful for the study of potentially important, yet poorly studied, community-level or cultural influences on breast cancer occurrence. A Wisconsin study , for insAPC = annual percent change; BMI = body mass index; CHIS = California Health Interview Survey; ER = estrogen receptor; HL = Hodgkin lymphoma; HT = hormone therapy; ICD-O-3 = International Classification of Diseases-Oncology, 3rd edition; NCI = National Cancer Institute; SEER = Surveillance, Epidemiology, and End Results; SFBA = San Francisco Bay area; WHI = Women's Health Initiative.CC has served as an expert witness for plaintiffs in hormone therapy tort litigation. The other authors declare that they have no competing interests.MW, EJ, SG, and PH-R contributed control data from the Marin County Breast Cancer Study of Adolescent Factors, the San Francisco Bay Area Breast Cancer Study, the Reproductive Factors in Hodgkin Lymphoma in Women study, and the Bay Area Thyroid Cancer Study, respectively, and participated in the interpretation of data and critical review of the manuscript. EC participated in the interpretation of data and drafting and critical review of the manuscript. TK and CC created the study design, conducted the acquisition, interpretation, and analysis of data, and led the writing and review of the manuscript. All authors read and approved the final manuscript.


Previous work has shown that extensive mammographic dysplasia in women aged less than 50 was strongly associated with breast cancer but that the radiological appearance of ductal prominence was not associated with risk. In the present paper we examine the association between these mammographic signs in the breast and histological patterns in the terminal ductal lobular unit (TDLU), the region of the breast where breast cancer is believed to originate. Surgical biopsies from a consecutive series of women aged less than 50 were reviewed and classified according to the histopathology of the epithelium in the TDLU. Mammograms from the same subjects were independently classified according to the extent of the radiological signs of dysplasia and ductal prominence. Degree of histopathology and the extent of mammographic dysplasia were associated and atypia of the ductal type was found more frequently in patients with extensive dysplasia. However, the strength and statistical significance of the association varied according to the radiologist who classified the mammograms. No association was found between degree of histopathology and ductal prominence. These results add to the evidence that extensive mammographic dysplasia in women aged less than 50 is a risk factor for breast cancer. They do not indicate that the radiological signs of dysplasia are caused by histological changes in the TDLU.


Characteristics probably associated with the fetal hormonal milieu have recently been shown to increase  or decrease (pregnancy toxaemia) breast cancer risk in the female offspring. However, it is unknown whether differences in pregnancy hormone levels may contribute to the marked geographical variation in breast cancer incidence. We have compared, in a highly standardized manner, pregnancy hormone levels in a population with high incidence and one with low incidence of breast cancer. Three hundred and four pregnant Caucasian women in Boston and 334 pregnant Chinese women in Shanghai were enrolled from March 1994 to October 1995. Levels of oestradiol, oestriol, prolactin, progesterone, human growth hormone, albumin and sex hormone-binding globulin were measured in maternal blood at weeks 16 and 27 of gestation and compared between the two study sites using non-parametric Wilcoxon's rank-sum test. Demographical, anthropometrical and pregnancy characteristics were ascertained through interview, and relevant variables concerning delivery and the newborn were abstracted from medical records and paediatric charts. During the first visit, median serum levels of all studied hormones were statistically significant, and in most instances substantially, higher among Chinese women, who have a low incidence of breast cancer, compared with American women, who have a high incidence of breast cancer. An analogous pattern was evident during the second visit, although the relative differences tended to be smaller. Further research is needed to identify lifestyle or other exogenous determinants of pregnancy hormone levels, as well as possible mechanisms by which they may influence carcinogenic processes in the breast and possibly other organs. © 1999 Cancer Research Campaign


Vitamin D status measured during adulthood has been inversely associated with breast cancer risk in some, but not all, studies. Vitamin D has been hypothesized to prevent breast cancer through genomic and non-genomic actions in cell-cycle regulation.A subset  of female participants from the prospective Cancer Prevention Study-II (CPS-II) Nutrition Cohort provided a blood sample from 1998-2001 and were followed through 2005. We measured serum 25-hydroxyvitamin D (25(OH)D) in 516 verified incident cases and 516 controls, matched on birth date (± 6 months), date of blood draw (± 6 months) and race. Information on medical history, risk factors and lifestyle was available from repeated questionnaires. We computed multi-variable odds ratios (OR) and 95% confidence intervals (95% CI) for the association between 25(OH)D quintile and breast cancer risk using unconditional logistic regression, controlling for matching factors and additional confounders.P = 0.60) for the top vs bottom quintile. Using a priori cut-points, the OR was 0.86 (95% CI 0.59-1.26), for ≥75 vs <50 nmol/L. Results were not different when the first two years of follow-up were excluded, or in analyses stratified by season, latitude, BMI, postmenopausal hormone use, or by tumor grade or estrogen receptor status.We observed no association between 25(OH)D and breast cancer D and postmenopausal breast cancer. We cannot rule out an association with 25(OH)D status earlier in life. Although breast cancer death rates have declined in recent years, breast cancer remains the most commonly diagnosed cancer and the second leading cause of cancer death in women in the US. . Risk fa2D)), the natural ligand for the nuclear vitamin D receptor. Although three case-control studies reported an inverse relation between circulating 25(OH)D and breast cancer risk [Breast cancer rates are higher in areas of the US with lower overall ultraviolet B radiation exposure, raising the possibility that higher levels of vitamin D could reduce breast cancer risk . Vitamincer risk -10, resucer risk -13.If vitamin D status is associated with a lower risk of breast cancer, this would represent an easily modifiable risk factor. To contribute to the limited and inconsistent prospective data on this topic of important public health interest, we examined circulating 25(OH)D concentrations in a nested case-control study of over 500 postmenopausal breast cancer cases and their matched controls from the Cancer Prevention Study-II (CPS-II).Women in this analysis were participants in the CPS-II Nutrition Cohort, a prospective study of cancer incidence among men and women from 21 states in the US, established by the American Cancer Society in 1992 . At enroFrom June 1998 to June 2001, blood specimens were collected from 21,965 women in the Nutrition Cohort who lived near a major collaborating hospital or medical clinic in their community, as described in detail elsewhere . All parEligible cases included women who reported a new diagnosis of breast cancer on a biennial CPS-II Nutrition Cohort Survey between the date of their blood draw and 30 June, 2005 (n = 514) or who did not report an incident breast cancer but for whom fatal breast cancer was identified through linkage with the National Death Index  (n = 2).We initially identified a total of 535 breast cancers diagnosed after the date of blood draw. We excluded women who were pre- or peri-menopausal (n = 3) at baseline, those with no appropriate match (n = 1), women with one or less vial of serum (n = 14), and one woman with an extreme 25(OH)D level (more than three standard deviations above the mean). The analysis is based on 516 cases and on an equal number of matched controls, aged 47 to 85 years at baseline. The duration of follow-up between the date of blood draw and breast cancer diagnosis ranged from less than 1 month to 6.9 years.We documented the estrogen receptor (ER) and progesterone receptor (PR) status of the tumors from medical record reports and through state cancer registries, when available. Information on ER and PR status was available for 76% and 73% of cases, respectively.Each control was individually matched to a case on birth date (± 6 months), race/ethnicity , and date of blood collection (± 6 months). We randomly selected one postmenopausal female control who had provided a blood sample and was cancer free (except for non-melanoma skin cancer) at the date of the diagnosis of the case. For two cases of 'other' race, an appropriately matched control was not available, and they were matched to a Caucasian control.Laboratory analyses were conducted at Heartland Assays, Inc . Case-control pairs were analyzed in the same batch in no particular order. A direct, competitive chemiluminescence immunoassay using the DiaSorin LIAISON platform, as described in detail elsewhere , was useAn average of five quality control samples were included in each batch of 100 samples. These quality control samples were from four separate sources: a pooled sample of non-multivitamin users, a pooled sample of multivitamin users, and two individual CPS-II Nutrition Cohort participants.Coefficients of variation (CV) for duplicate serum aliquots included in all laboratory sample batches were calculated for each of the four quality control samples, and ranged from 6.4 to 8.8%. Intra-batch and inter-batch CVs were 5.4% and 8.1%, respectively.Information on breast cancer risk factors and potential confounders was ascertained from responses to biennial questionnaires from the CPS-II cohort. For the current analysis, we obtained information on current BMI, date of blood draw, and use of individual vitamin D supplements from the brief questionnaire completed at blood draw. Information on reproductive risk factors, history of benign breast disease, family history and education was assessed on earlier questionnaires. Other variables, including alcohol use, postmenopausal hormone use, diet, recreational physical activity and zip code (for latitude) were collected from the comprehensive 1999 questionnaire. For these variables, 24 cases (5%) completed the questionnaire after diagnosis of breast cancer. Sensitivity analyses removing these cases and their controls did not change the risk estimates.Circulating 25(OH)D (nmol/L) was normally distributed and was not improved with log transformation; thus, we used untransformed values. We divided participants into quintiles based on the distribution of serum 25(OH)D in the control group. We observed the expected differences in 25(OH)D concentrations by season of blood draw: mean levels (± standard deviation) were 49.1 (± 21.2) in Winter (December to February), 53.7 (± 25.1) in Spring (March to May), 58.1 (± 22.2) in Summer (June to August) and 59.5 (± 20.3) in Fall (September to November). For stratified analyses, we classified season according to two general seasons: a 'warm' season (June to November) and a 'cool' season (December to May). Independent correlates of 25(OH)D among controls in our study were evaluated using linear regression models.Our primary statistical approach used unconditional logistic regression to estimate the odds ratio (OR) and 95% confidence intervals (CI) for risk of cancer comparing quintiles 2 to 5 with the lowest quintile, to preserve statistical power when conducting stratified analyses on factors that were not matched by design. In sensitivity analyses, we also conducted conditional logistic regression on main effects models and results were similar. We first assessed the relation between serum 25(OH)D and breast cancer risk adjusting only for matching factors ).In addition to controlling for matching factors, we included as covariates those factors that were correlated with serum 25(OH)D concentrations and/or related to breast cancer risk in this sample. These included: season of serum collection ; weight change since age 18 years ; body mass index , 30.0+), and combinations of number of live births  and age of mother at first birth (<25 or ≥25). Adjustment for alcohol consumption ; recreational physical activity in metabolic equivalents (MET) hours per week (quartiles); family history of breast cancer ; history of benign breast disease ; age at menarche , >13 years); age at menopause , and postmenopausal hormone use  did not change the effect estimates and therefore were not included. To test for trend, we created a continuous variable consisting of the median 25(OH)D value from each quintile, and included this variable in the regression model.a priori cut-points of less than 50 nmol/L as deficiency (referent), 50 to less than 75 nmol/L as 'insufficient', and 75+ nmol/L as 'adequate' [We additionally examined the association between 25(OH)D and breast cancer risk using dequate' . We examdequate' . The lin2), time of year of blood draw ('warm' and 'cool' seasons), latitude (above and ≤37°), tertile of total calcium intake, multivitamin supplement use (assessed in 1999) and use of individual vitamin D supplements  reported at blood draw. Interaction terms were cross-product terms of the tertile measure  and dummy variables for potential effect modifiers. Multiplicative interactions were evaluated using the likelihood ratio test.We conducted stratified analyses comparing breast cancer risk across tertiles of 25(OH)D concentration within strata of postmenopausal hormone use (never and ever use), BMI . All statistical analyses were conducted using SASv.9.2 .We conducted sensitivity analyses excluding cases diagnosed during the first two years of follow-up, and using season-specific 25(OH)D quintile cut-points . We alsoCharacteristics of the postmenopausal breast cancer cases and controls are provided in Table Several factors predicted circulating 25(OH)D among controls (results not shown). Winter season predicted significantly lower 25(OH)D concentrations compared with Summer. Current BMI and weight gain each individually predicted 25(OH)D, but when considered simultaneously, current BMI was the stronger predictor. Recreational physical activity was associated with significantly higher concentrations, likely due to greater sunlight exposure during outdoor exercise. Total vitamin D intake  was a statistically significant predictor of 25(OH)D in univariate and multivariate models. An increment of 800 IU vitamin D corresponded to a 16 nmol/L higher circulating 25(OH)D in both models (not shown).P = 0.6). A cubic spline graph . Results were similar when 227 cases diagnosed in the first two years after blood draw were excluded . Results using season-specific 25(OH)D cut-points were also similar . In sub-analyses using absolute cut-points of 75 or more vs less than 50 nmol/L, the ORs remained null for all cases .Table h Figure  illustrain situ, localized, or regional/distant. In sub-analyses using absolute cut-points of 75 or more vs less than 50 nmol/L; ORs (± 95 CI) were: ER+, OR = 0.86, 95% CI 0.56 to 1.32, P = 0.8; ER-, OR = 0.90, 95% CI 0.35 to 2.30, P = 1.0; in situ, OR = 0.64, 95% CI 0.32 to 1.30, P = 0.3; localized, OR = 1.16, 95% CI 0.76 to 1.78, P = 0.3; regional and distant, OR = 0.39, 95% CI 0.16 to 0.92, P = 0.08. Thus, when using established cut-points, individuals who had sufficient vs. deficient levels were at borderline lower risk for regional/distant tumors. When modeled by each 10 nmol/L change in 25(OH)D, the multivariate-adjusted ORs and 95% CIs were as follows: all cases ; ER+ cases ; ER- cases ; in situ cases ; localized cases  and regional/distant cases .Associations by ER status and tumor subtype were examined using tertiles of 25(OH)D concentration to allow comparison among groups with smaller numbers (Table P interaction = 0.003). The relation did not vary by use of vitamin D-containing multivitamin supplements (measured in 1999), but a significant interaction was observed for use of individual vitamin D supplements  at blood draw: among women who reported not using vitamin D supplements, the OR for the top vs bottom tertile was 0.82 (95% CI 0.53 to 1.27), whereas among current users it was 1.82 .The association between 25(OH)D and breast cancer risk did not vary by history of postmenopausal hormonal therapy, BMI, weight gain, 'warm' and 'cool' season of blood draw, or calcium intake. A significant interaction was observed by latitude: the OR among women living at 37° or more latitude who were in the top tertile of 25(OH)D (64 nmol/L+) vs bottom tertile (<46 nmol/L) was 0.35 (95% CI 0.17 to 0.73), vs OR = 1.07 (95% CI 0.59 to 1.95) among women living at <37° D concentrations assessed up to six years prior to diagnosis were not associated with risk of postmenopausal breast cancer. This finding did not change when we excluded cases occurring in the first two years after blood draw, and were not significantly different in analyses stratified by BMI, postmenopausal hormone use, season of blood draw, calcium intake, or multivitamin use. Likewise, 25(OH)D concentrations were not significantly associated with cancer subgroups defined on the basis of ER/PR status, or tumor stage/grade.P = 0.06). In contrast, among 1005 postmenopausal cases and 1005 controls from the Prostate, Lung, Colorectal and Ovarian Cancer Screening Trial [P trend of continuous 25(OH)D = 0.76) [Our findings contribute to the relatively few prospective studies published on this topic -13. In ang Trial  reported = 0.76) . All thr = 0.76) -10 repor = 0.76) ,22. A se = 0.76) , but thea priori cut-points for deficiency, insufficiency and sufficiency [Our study did not indicate the existence of a threshold for a protective association with breast cancer with circulating 25(OH)D, which has been hypothesized by some . In the ficiency , which rBody fat and weight gain were moderate negative confounders in our study (with the net effect of raising the OR when added to either the age-adjusted model or multivariate models), and BMI independently predicted circulating 25(OH)D. Body fat is a large storage reservoir for 25-OH-vitamin D, and lower circulating 25(OH)D in obese individuals is thought to be related to greater uptake into fat tissues, rather than lower sun exposure or less effective vitamin D synthesis . In the P interaction = 0.047). Reasons for this apparent interaction are unclear. It is possible that the source of vitamin D may be important. For example, at any given level of serum 25(OH)D, non-supplement users obtain a higher proportion of their vitamin D from cutaneous synthesis, and/or greater intake of fortified milk or fish than do supplement users. Vitamin D in the diet is strongly correlated with dietary calcium, which also may play a role in preventing breast carcinogenesis [We found that 25(OH)D-breast cancer associations were modified by individual vitamin D supplement use at blood draw, with an inverse association among non-current vitamin D supplement users and a positive association among current supplement users D did not correctly classify women according to their usual, year-round vitamin D status. Circulating 25(OH)D reflects exposure to sun and diet from approximately the prior two months. Further, 25(OH)D fluctuates according to season of the year. To minimize this extraneous source of variation, we controlled for season of blood draw, and in sensitivity analyses, we used season-specific cut-points. Analyses stratified by season of blood draw were not statistically different.Our findings do not support an association between adulthood 25(OH)D status and postmenopausal breast cancer risk. Future studies with the ability to examine a wider distribution of 25(OH)D, vitamin D status earlier in life, and with greater statistical power to examine tumor subtypes and stage would add importantly to the literature.2D: 1,25-dihydroxyvitamin D; 25(OH)D: 25-hydroxyvitamin D; BMI: body mass index; CPS-II: Cancer Prevention Study-II; CI: confidence interval; CV: coefficient of variation; ER: estrogen receptor; METS: Metabolic equivalent of task; OR: odds ratio; PR: progesterone receptor; SD: standard deviation.1,25(OH)The authors declare that they have no competing interests.MLM conceived of the study, participated in the design, and drafted the manuscript. VLS participated in the design, coordinated laboratory sample processing, and contributed to drafting the manuscript. RP participated in study design and conducted statistical analyses. EBB participated in study design and processed laboratory samples. EJJ participated in the design and contributed to drafting the manuscript. RLH conducted laboratory analyses and contributed to sections of manuscript draft. SMG participated in the design and contributed to drafting the manuscript. MJT contributed to drafting the manuscript. EEC participated in the design of the study. All authors read and approved the final manuscript.


Breast cancer is the most commonly diagnosed cancer among women in many Western populations , althougThese international variations in incidence rates coupled with the results from migration studies suggest breast cancer risk is influenced by environmental exposures, reproductive history, and lifestyle choices, including diet , was not associated with breast cancer risk. In all, these exciting findings by Wu et al underscore the importance of considering exposure dose and timing when evaluating the effects of soy intake on breast cancer risk.One of the most important contributions of the meta-analysis undertaken by Wu et al sheds light on these issues. For example, many of the investigations conducted to date have used the case–control study design, which may yield biased effect estimates due to possible differential recall of exposure based on the subjects' case–control status. However, as shown by Wu et al, results among Asian populations from both cohort and case–control studies yielded nearly identical summary effect estimates. Further, if recall bias was the underlying reason for the inverse association noted in the meta-analysis presented by Wu et al, then the risk reduction associated with isoflavone intake should have been evident among case–control studies conducted among Asian and Western populations, which was not the case. Thus, recall bias does not appear to be the reason for the observed risk reductions associated with soy intake among Asian populations.Other concerns have also contributed to the difficulty in interpreting the numerous population studies on the soy–breast cancer association, and again the thoughtful methodologic approach of Wu An important issue that still remains unclear is whether any of the biologically active anti-cancer flavonoid compounds influence breast cancer risk among Western populations. Most previous epidemiologic studies have been hampered by focusing solely on intake of isoflavones, which are usually consumed by Westerners through intake of soy-based products such as processed foods. Challenges associated with assessing intake of isoflavones from these types of products could perhaps be overcome by utilising a biomarker of exposure, although single measures common to epidemiologic studies are not easily interpreted. Instead, it may be more productive to consider intake of a broader range of flavonoids, many of which occur in foods more frequently consumed by Western populations (such as tomatoes, green peppers, berries, and citrus fruits (et al also highlights other understudied soy–breast cancer issues. In particular, the authors note that few investigators have formally evaluated the effects of soy intake on survival among women diagnosed with breast cancer. For example, one study conducted among a cohort of women with breast cancer in Shanghai (The review by Wu et al notably demonstrate that relatively high intake of 10–20 mg per day of isoflavones, a consumption level frequently found among Asian populations, is inversely associated with a 10–30% reduction in risk of breast cancer. Lower isoflavone intake of <1 mg, a level more commonly consumed by Western populations, does not appear to have any beneficial effects on breast cancer risk. Additional research is needed to assess whether soy consumption that begins during the adult years is sufficient to decrease risk, or whether only intake patterns that were initiated during childhood affect the subsequent risk of developing breast cancer. In addition, clarification is needed on whether intake of other flavonoid classes, particularly those in foods such as fruits and vegetables more frequently consumed by Western populations, are associated with breast cancer risk reductions as has been shown for flavones in three recently published studies. Lastly, whether consumption of isoflavones or other flavonoid classes affects mortality among breast cancer survivors has been evaluated in only two previous reports, with conflicting results, and thus this issue also deserves to be more extensively evaluated.In sum, the meta-analysis by Wu


Four hundred and fifteen males and 367 females who had invasive malignant tumours of the salivary glands as their first cancer diagnosed in Connecticut between 1935 and 1978 were identified and followed 2342 and 2868 person-years respectively. Overall a slight excess of second primary cancers (relative risk 1.35) was observed. Significant excesses were noted for respiratory cancers in males (relative risk 2.8) and for ovarian cancer (relative risk 5.3) but not breast cancer (relative risk 1.3) in women. Possible reasons for excesses at these sites are discussed, but it seems most likely they are related to small number variation.


Several studies have examined the effect of genetic variants in genes involved in the estrogen metabolic pathway on mammographic density, but the number of loci studied and the sample sizes evaluated have been small and pathways have not been evaluated comprehensively. In this study, we evaluate the association between mammographic density and genetic variants of the estrogen metabolic pathway.A total of 239 SNPs in 34 estrogen metabolic genes were studied in 1,731 Swedish women who participated in a breast cancer case-control study, of which 891 were cases and 840 were controls. Film mammograms of the medio-lateral oblique view were digitalized and the software Cumulus was used for computer-assisted semi-automated thresholding of mammographic density. Generalized linear models controlling for possible confounders were used to evaluate the effects of SNPs on mammographic density. Results found to be nominally significant were examined in two independent populations. The admixture maximum likelihood-based global test was performed to evaluate the cumulative effect from multiple SNPs within the whole metabolic pathway and three subpathways for androgen synthesis, androgen-to-estrogen conversion and estrogen removal.Genetic variants of genes involved in estrogen metabolism exhibited no appreciable effect on mammographic density. None of the nominally significant findings were validated. In addition, global analyses on the overall estrogen metabolic pathway and its subpathways did not yield statistically significant results.Overall, there is no conclusive evidence that genetic variants in genes involved in the estrogen metabolic pathway are associated with mammographic density in postmenopausal women. Mammographic breast density is one of the strongest risk factors for breast cancer. Several studies have shown that women with extensive dense tissue are at two to six times higher risk of developing the disease than women of similar age with lower mammographic density ,2. A strMammographic density is strongly correlated with hormone exposure profiles of women . Severalet al., manuscript submitted). No significant SNP association was evident after correction for multiple testing, but pathway-based global tests revealed significant association evidence for the overall estrogen metabolic pathway (P = 0.034) and, in particular, the androgen-to-estrogen conversion subpathway (P = 0.007). In the present study, we comprehensively examine genetic variation in the estrogen metabolic pathway with mammographic density. The number of SNPs and genes studied provides the most extensive coverage to date with respect to studying mammographic breast density.We recently reported the results of a study evaluating a total of 239 SNPs in 34 estrogen metabolic genes in 1,596 breast cancer cases and 1,730 population controls from Sweden, of which the outcome variable was breast cancer  and six other ethical review boards in the respective regions in which the subjects were based, and informed consent was obtained from each participant.Validation of SNPs with significant associations was performed using mammographic density data from two other studies.The process of collecting mammographic density data in this study has been described previously . Film maet al. (manuscript submitted). A total of 1,007 SNPs were selected from 35 genes and their 30 kb flanking sequences that code the enzymes involved in estradiol or estrone metabolism and are expressed in the breast. These SNPs were genotyped in 92 Swedish control samples to assess linkage disequilibrium patterns, to select tagging SNPs (tagSNPs) and to evaluate their coverage.The process of gene and SNP selection has been described in detail by Low tagSNPs program [R2 coefficient, which quantifies how well the tagSNP haplotypes predict the genotype or the number of copies of haplotypes an individual carries. The performance of tagSNPs in capturing unobserved SNPs within the genes was evaluated using a SNP-dropping analysis. In brief, each of the genotyped SNPs was dropped in turn and then tagSNPs were selected from the remaining SNPs so that their haplotypes predicted the remaining SNPs with an R2 value of 0.85. In total, 312 tagSNPs from the 35 genes were selected for genotyping.Haplotypes were reconstructed using the partition-ligation-expectation-maximization algorithm  implemen program . A subseFigure P < 0.05/312) were excluded from further analysis. The genotype concordance was >99%, suggesting high genotyping accuracy. Overall, 239 tagSNPs from the 34 genes were successfully genotyped and used in statistical analysis.DNA was extracted from 4 ml whole blood using the QIAamp DNA Blood Maxi Kit  according to the manufacturer's instructions and nonmalignant cells in paraffin-embedded tissue using a standard phenol/chloroform/isoamyl alcohol protocol. Genotyping was performed using the primer extension-based assay from Sequenom  according to the manufacturers' instructions. DNA samples were randomly assigned to the plates carrying positive and negative controls, and all genotyping results were generated and checked by laboratory staff unaware of the case-control status. SNPs with a call rate <85%, minor allele frequency <1% or out of Hardy-Weinberg equilibrium  [A likelihood ratio test was performed for each SNP. Normal quantile-quantile plots were used to examine the distributions of the -logbal test  was perf(v2.8.0)  and admi(v2.8.0)  were useSNP associations with mammographic density were validated in 1,590 women genotyped with the Illumina HumanHap500 as part of the Cancer Genetic Markers of Susceptibility Project (CGEMS) . The CGEWe collected mammograms as close as possible to the date of blood collection (1989 to 1990). To assess mammographic density, the craniocaudal (CC) views of both breasts were digitized at 261 μm/pixel with a Lumysis 85 laser film scanner, which covers a range of 0 to 4.0 optical density. The software for computer-assisted thresholding was developed at the University of Toronto . We usedThe second validation population consisted of a set of controls from an ongoing breast cancer case-control study at the Mayo Clinic. Briefly, the Mayo Clinic Breast Cancer Study is an Institutional Review Board-approved, clinic-based, case-control study initiated in February 2001 at Mayo Clinic, Rochester, MN, USA. The study design has been presented previously ,31. ClinThe present analysis genotyped Caucasian controls (99% of study participants) enrolled through September 2007, who had mammograms available, representing 995 total controls , of which 783 were postmenopausal. Screening mammograms were ascertained close to the enrollment date and the left CC view was digitized on an Array 2905HD Laser Film Digitizer, which covers a range of 0 to 4.7 optical density. Percentage mammographic density was estimated by an expert reader  on the lP = 0.0126), parity (P < 0.0001), family history of breast cancer (P = 0.0002) and percentage density (P = 0.0017). Cases were found to have higher percentage density (mean ± standard deviation: 16.7 ± 14.3) than controls (14.6 ± 14.0). No significant difference was found for age, height, weight, body mass index, age at menarche, age at menopause or HRT usage.Our dataset consisted of 1,731 postmenopausal women, of which 981 were breast cancer cases and 840 were controls .Table S4 in Additional file y Figure . A seconP = 0.5268). Pathway-based multi-SNP association analyses revealed no significant association between percentage density and genetic variations in the overall estrogen metabolic pathway, or any of the related subpathways , HRT, and menopausal status, the estrogen metabolic pathway has been a candidate pathway for the search of genetic variants related to mammographic density. Most of the variants in the candidate breast cancer genes evaluated in previous studies, however, have been concluded to be only weak predictors of mammographic density . AssociaIn our study, no correlation was found between the estimates of SNP effects on breast cancer risk and mammographic density, suggesting that the same SNPs associated with breast cancer risk are not directly correlated with mammographic density. Tamimi and colleagues reported that mammographic density and circulating sex steroid levels were independently associated with breast cancer risk in postmenopausal women . In addiMany studies examining the effects of exogenous estrogen exposure are in agreement with the view that estrogen has limited effects on mammographic density. Very often, the combined estrogen plus progestin regimen was found to affect mammographic density more than the estrogen-only regimen -41, suggStrengths of the present study include the large sample size and extensive coverage of SNPs in the estrogen metabolic pathway. In a review by Kelemen and colleagues, the authors summarized that previous genetic association studies exploring the relationship between the estrogen metabolic pathway and mammographic density had sample sizes ranging from between 232 and 1,260 women [A limitation of the present work is that it includes different mammogram views across the different studies. The main study on Swedish women utilized the medio-lateral oblique view, while mammograms of the CGEMS and of the Mayo Clinic were taken using the CC view. Several studies, however, have shown correlation of densities from the medio-lateral oblique and CC views ,47, and As mammographic density is generally considered an intermediate phenotype for breast cancer, the identification of genes that influence mammographic density would play an important role in risk prediction of breast cancer prior to the start of mammography screenings and shed light on the mechanisms behind breast cancer carcinogenesis. Overall, there is no conclusive evidence that genetic variants in genes involved in the estrogen metabolic pathway are associated with mammographic density in postmenopausal women. This knowledge will be helpful for directing the focus of future studies to alternative pathways that may be responsible for a larger bulk of the genetic component of mammographic density.CC: craniocaudal; CGEMS: Cancer Genetic Markers of Susceptibility Project; HRT: hormone replacement therapy; SNP: single nucleotide polymorphism; tagSNP: tagging single nucleotide polymorphism.The authors declare that they have no competing interests.JLi participated in the study design, carried out the analyses and drafted the manuscript. LE digitized and obtained readings for the mammograms. RMT, SL, DJH, CMV, FJC and CGS contributed to the validation of this study. PL coordinated the Innovator project which contributed data on birthweight and mammographic density. JLi, KH, KC, JLiu and PH conceived of the study, and participated in its design and coordination and helped to draft the manuscript. All authors read and approved the final manuscript.Tables S1 to S3. Table S1 presents a list of SNPs in the androgen synthesis subpathway and their corresponding regression coefficients and likelihood ratio test P values. Table S2 presents a list of SNPs in the androgen to estrogen conversion subpathway and their corresponding regression coefficients and likelihood ratio test P values. Table S3 presents a list of SNPs in the estrogen removal subpathway and their corresponding regression coefficients and likelihood ratio test P values.Click here for fileTable S4. Table S4 presents genes containing polymorphisms within the estrogen metabolic pathway evaluated in relation to mammographic density.Click here for fileTable S5. Table S5 presents validation results of significantly associated SNPs in the Nurses' Health Study (NHS) and the Mayo Clinic Breast Cancer Study (MBCS).Click here for file


Sir,vs a low-risk case at the time of diagnosis by socioeconomic position. This we investigated in a nation-wide population-based cohort of 28 765 women diagnosed with breast cancer in Denmark between 1983 and 1999 using a wide range of socioeconomic factors such as educational level and income (We thank von Euler-Chelpin and Lynge for their interest and comments. However, we believe their comments result from some misunderstandings. First of all, our study was not a case–control study. Secondly, and more importantly, we did not aim to investigate breast cancer incidence by socioeconomic level and this could not have been estimated from our study. Our aim was to assess, given a breast cancer, the probability of having a high-risk In Denmark, as in most other Western countries, the overall incidence of breast cancer increases with higher socioeconomic position . We have


Previous results have suggested an association of the region of 19q13.3 with several forms of cancer. In the present study, we investigated 27 public markers within a previously identified 69 kb stretch of chromosome 19q for association with breast cancer by using linkage disequilibrium mapping. The study groups included 434 postmenopausal breast cancer cases and an identical number of individually matched controls.RAI  and the 5' portion of XPD to be associated with this cancer. The region corresponds to a haplotype block, in which there seems to be very limited recombination in the Danish population. Studying combinations of markers, we found that two to four neighboring markers gave the most consistent and strongest result. The haplotypes with strongest association with cancers were located in the gene RAI and just 3' to the gene. Coinciding peaks were seen in the region of RAI in groups of women of different age.Studying one marker at a time, we found a region spanning the gene RAI-3'd1, located downstream of the transcribed region of RAI, which was more strongly associated with breast cancer than any other marker we have tested , p = 0.01, cases up to 55 years of age).In a follow-up to these results we sequenced 10 cases and 10 controls in a 44 kb region spanning the peaks of association. This revealed 106 polymorphisms, many of which were not in the public databases. We tested an additional 44 of these for association with disease and found a new tandem repeat marker, called RAI-3'd1 to be (part of) the cause for the association of the chromosome 19q13.3 region's association with cancer.We expect the marker  XPD  and ERCC1, one gene presumably relating to apoptosis, RAI , and one presumably involved in ribosomal RNA transcription, ASE1 alias CD3EAP  we tested an additional 44 polymorphisms for association with disease and identified a 5-bases tandem repeat polymorphism, NT_011109.15 g.18147012dupATTTT(2_11), here called RAI-3'd1, which discriminates cases from controls better than any previously investigated polymorphism in this region, and which may well be the effector causing this haplotype block to associate with cancer.In the present study, involving 434 breast cancer cases and an identical number of controls, we initially investigated 27 public markers within the 69 kb stretch for association with breast cancer, using linkage disequilibrium mapping based on single markers, as well as on haplotypes combining several neighboring markers. We found that the region formed two haplotype blocks, and that associations were concentrated on the gene This investigation deals with breast cancer in women. The cohort consisted of 24 697 Danish postmenopausal women age 50 – 64 at inclusion in 1993 – 1997. 434 of these women developed breast cancer before 2001. The cases and a set of individually matched controls (see Methods for details) were analyzed as a nested case-control study. [RAI intron8-3 denotes the third SNP investigated in the eighths intron of RAI. In the relative few polymorphisms that were in/dels or repeat polymorphisms, the serial number was preceded by a d, for instance RAI-3'd1. The markers genotyped in this study are listed in Table We investigated DNA polymorphisms in a 69 kb sub-region of 19q13.3 for association with breast cancer. To improve readability each polymorphism investigated was given a trivial name in accordance with its positions relative to the nearest gene followed by a hyphen and a serial number, for instance a priori assumptions about the distribution of genotypes are needed. Figure RAI and the 5' region of the gene XPD. The singular values for the polymorphism RAI-3'd1 will be discussed later. The corresponding curves for women aged less than 55 showed an even higher relative risk in the same region with a peak value of RR = 6.25 (1.72 – 20), p(trend) = 0.005 (results not shown).Next, we calculated the relative risks of breast cancer and the corresponding confidence limits for each marker comparing the variant homozygotes to homozygotes for the wildtype allele. However, if the calculation gave an RR value less than 1, we used the reciprocal value of the RR and the reciprocal value of the upper confidence limit instead. This analysis has the advantage that no RAI-3'd1, to be described later.We also used the analytical technique known as haplotype trend regression , more spXPD exon6 to RAI intron8-3, corresponding to the region with association to cancer, whereas the other haplotype block spanned 6 markers and 9 kb from RAI intron1-1 to ASE1 exon3-3.Finally, we calculated the pair-wise linkage disequilibrium of all markers in the controls Figure . The regBased on all these results we chose to sequence PCR fragments spanning the region from relative position 13 to 59 kb in breast cancer cases and controls. Initially, we selected 10 cases and 10 controls for analysis. The actual number of reliable sequences obtained varied somewhat from one part of the region to the other. Table 9 lists the 106 detected polymorphisms including their positions, their rs-numbers if assigned, and the surrounding sequences [See Additional File Based on the sequencing results, we analyzed 44 new polymorphisms using ABI3100, Lightcycler and Sequenom ) than other markers in the youngest age group and showed a clearer tendency in the other age groups  than any we had seen .To better understand the nature of the association, we tabulated the frequencies of individual haplotypes made up of the 3 neighbouring markers centring on the position at 24 kb with maximal evidence of association with breast cancer . We found the persons that were RAI-3'd1SS to have an increased risk of breast cancer between 50 and 60 years of age .We used several ways to try to locate the effector(s). The overall conclusion was clear: One effector is located in the haplotype block between the markers RAI-3'd1, that showed even stronger association with cancer (p = 0.0008). Bonferroni adjustment for the multiplicity of testing indicates that with an average correlation of 0.32 among a total of 68 polymorphisms, any p-value below 0.003 will make the whole set significant [We resequenced the region of strongest association identified by haplotype trend regression. Here we discovered a new 5-base tandem repeat polymorphism, nificant ). Thus, RAI-3'd1 may well be the effector. The haplotype data of Table RAI-3'd1S were associated with increased risk of cancer, irrespective of the surrounding polymorphisms. We would not expect this result if RAI-3'd1 only was a correlate of the effector. However, mapping inside the haplotype block with the use of data from the present population is not very sensitive. For instance, the markers XPD-5'2 and RAI intron11, spaced 12 kb apart, were synonymous in 799 of 813 persons. Studies of this polymorphism in populations of different ethnicity could prove very useful, but functional assays will be necessary before a final conclusion can be reached.The polymorphism RAI region with that caused by the BRCA genes. The RAI region is either recessive or suffers from strong haplo-insufficiency, and it roughly doubles the risk of breast cancer, thus the penetrance is low. The BRCA genes on the other hand are dominant, and they increase the breast cancer risk about 8 fold, i.e. penetrance is high. However, the risk allele of RAI is very common. In our study approximately 40 percent of the controls were homozygotes for the risk allele. A conservative estimate of the ethiological fraction suggests that at least 20 percent of breast cancers is caused by RAI-3'd1, and the number may be twice as high. Thus in terms of the number of cancers caused in the Danish population, RAI seems to exceed the BRCA genes in importance.It may be useful to compare aspects of the breast cancer risk caused by the RAI-3'd1S, occurred with different neighboring SNP alleles at reasonable frequencies. In contrast the protective allele RAI-3'd1L almost exclusively occurred together with RAI-3'd3S and RAI-3'4C. Further investigation has shown that this tendency was reproduced with a number of the surrounding SNPs. Generally, the less frequent haplotypes involving RAI-3'd1S were approximately 10 fold more common than the less frequent haplotypes involving RAI-3'd1L were. This was true for 12 out of 14 SNPs in the 23 kb interval from XPD exon6 to RAI intron8-1. The exceptions were RAI-e13 and RAI-3'7. In accordance, combined analysis of the 12 SNPs using HelixTree to form a 12 point haplotype indicated that 13 less frequent haplotypes involving RAI-3'd1S constituted 23 percent of the chromosomes, corresponding to 56 percent of the major haplotype with RAI-3'd1S. In contrast, 7 less frequent alleles involving RAI-3'd1L only constituted 2 percent of the chromosomes, corresponding to only 8 percent of the major haplotype with RAI-3'd1L (results not shown). Thus, the haplotypes containing RAI-3'd1S seem more diverse than their counterparts containing RAI-3'd1L do. The simple interpretation of this phenomenon would be that RAI-3'd1S is the ancestral allele in Caucasians.From Table RAI-3'd1 influences the expression of RAI. Our hypothesis is that RAI-3'd1 modulates a silencer or enhancer. The influence of RAI-3'd1 may not be limited to RAI. It is a curious fact that this short stretch of chromosome 19, comprising only 69 kb, contains two DNA repair genes and one apoptosis gene. One could imagine some sort of higher-order control common for this cluster of related genes. Studies of the expression of XPD, RAI and ERCC1 in lymphocytes from humans have indeed shown that their mRNA levels are tightly coordinated [There are at present no known polymorphisms affecting the amino acid sequence of the RAI protein. Presumably, rdinated ,19.RAI-3'd1 is modified RAI expression, one result should be modified apoptosis. RAI protein is an inhibitor of RelA, a subunit of the transcription factor NF-κB [RelA and c-rel [RelA product, RAI protein would presumably shift this balance towards apoptosis. In accordance with this view, we have found that RAI mRNA increases dramatically when cells undergo apoptosis. We have also found that simultaneous transfection with siRNAs destabilizing RAI mRNA reduces apoptosis in non-transformed cells. Thus, we believe that RAI expression is instrumental for apoptosis [If a primary effect of or NF-κB ,21. NF-κnd c-rel . By neutpoptosis . Howeverpoptosis . For inspoptosis . FurtherThe previous reports, that have associated our region of interest with the risk of getting basal cell carcinoma, melanoma, lung cancer, glioma, breast cancer, bladder cancer, and possibly head and neck cancer -10,26,27RAI-3'd1 may be the cause for the extra risk for breast cancer. We think it may modulate an enhancer or silencer influencing RAI and possibly other genes.We found the region around the RAI gene to be strongly associated with breast cancer, in particular during the early menopause. The marker This study was performed in compliance with the Helsinki declaration. The project was approved by the Science Ethical Committee of Copenhagen (j.nr: 01-345/93/(KF) 11-037/01/11-124/01). Written and verbal informed consent was obtained from all participants."Diet, Cancer and Health" is a Danish prospective follow-up study . IndividThe study sample for breast cancer included the 29 549 women recruited within "Diet, Health and Cancer". Eight women were excluded from the study because they did not fill in the lifestyle questionnaire. Because the present analysis aimed at the subgroup of women who were post-menopausal at study entry, we excluded 4844 supposedly pre-menopausal women, including 4798 women who had reported at least one menstruation no more than 12 months before entry and no use of HRT, nine women who gave a lifetime history of no menstruations, and 37 women who did not answer the questions about current or previous use of HRT, leaving 24 697 postmenopausal women.Each cohort member was followed up for breast cancer occurrence from the date of entry, i.e. date of visit to the study centre until the date of diagnosis of any cancer (except for non-melanoma skin cancer), date of death, date of emigration, or 31 December 2000, whichever came first. A total of 434 women were diagnosed with incident breast cancer during the follow up period. Controls were selected among the remaining cohort participants using a nested case-control approach. One control was selected for each of the 434 cases. The control was cancer-free at the exact age at diagnosis of the case and was further matched on age at inclusion into the cohort , certainty of postmenopausal status , and use of HRT at inclusion into the cohort (current/former/newer). Of the 434 pairs , 19 pairs were excluded due to lack of blood sample in either case or control, leaving 415 pairs for study. Additional pairs were excluded due to problems with the genotyping procedure in one or both members of the pair..Resequencing was performed on PCR fragments from DNA of persons selected for different haplotypes among the known markers in the region. 7 of the cases were further selected for having gotten breast cancer before age 55. The DNA sequences to be amplified and resequenced were obtained from databases. Repetitive sequences were determined and masked using RepeatMasker software. Primers were chosen using Primer3 software. PCR amplification was performed in a 10 μL volume and the product was purified with Centricon 100 (Ambion). Each sequencing reaction contained 2.5 μL of the purified PCR product, 6.5 μL of water, 2.0 μL of BigDye version 3 mix (Applied Biosystems), and 1.0 μL of 5× sequencing buffer, according to the protocol of the dye manufacturer. Electrophoresis and sequencing detection were performed using an ABI PRISM 3730 × l DNA Analyzer (Applied Biosystems).Table Data recording, calculations and tests of allele frequencies were performed in SPSS and Excel. Calculation of the relative risk and confidence intervals for the single polymorphisms was done using the procedure phrec in the program package SAS . Simultaneous analysis of multiple SNPs employing haplotype trend regression  was perfThe authors declare that they have no competing interests.BAN provided concepts and ideas, performed part of the data analysis and drafted the manuscript. UV contributed to the concepts and performed part of the marker analyses. AO performed part of the data analysis. MN, BH, ZB, ER, CK, MM, AK and ML established and performed most of the marker analyses. QZ performed the sequencing and part of the associated analysis. OR-N, SC, KO and AT established the cohorts. LB and ADB provided ideas and analyses.The pre-publication history for this paper can be accessed here:SNP typing and SNP identification. The conditions for the SNP typing and the description of the SNPs found by sequencing are stored as a .pdf file in the Additional File Click here for file