Women's awareness of ovarian cancer (OC) risks, their attitudes towards and beliefs about screening, together with misunderstandings or gaps in knowledge, may influence screening uptake.In total, 21 715 post-menopausal women completed questionnaires before randomisation into the UK Collaborative Trial of Ovarian Cancer Screening.In all, 42.3% correctly identified their lifetime risk of OC; 87.1% knew that a family history of OC increased risk, but only 26.7% appreciated the association with a family history of breast cancer. Although 38.2% acknowledged increased risk post-menopause, only 8.8% were aware that OC diagnoses are highest in women over 65 years. Few (13.7%) recognised the association between pregnancy and reduced OC risk or protective effects of breastfeeding (6.2%). There were common misconceptions; 37.2% thought that an abnormal cervical smear and 26.4% that oral contraception increased the likelihood of OC. Although 84.4% recognised that most ovarian masses are benign, 20.2% thought having had a benign cyst increased OC risk. Most (99.4%) believed that a high uptake of OC screening would reduce mortality and (96.2%) that screen-detected cancers would have an improved prognosis.The results show a need for improved public understanding about OC risks and provide important information for GPs and health educationalists about initiatives needed for future awareness, prevention and screening programmes. In 2007, 4317 UK women died from ovarian cancer (OC), the fifth most common cancer in women . EarlierLittle is known about the awareness of the risks of OC or views about screening among women in the general population of the United Kingdom, although a survey commissioned in 2007 by an OC charity and Dr Foster Intelligence suggested that 59% of women were unaware of any risk factors associated with OC . Most puIn this paper, we report the knowledge and pre-randomisation beliefs of 21 715 women participating in UKCTOCS regarding OC risk, perceived risk factors, their attitudes towards screening and the associations of all these with sociodemographic, psychological characteristics and medical history.The design of the main UKCTOCS has been described in detail . It has This paper describes data from pre-randomisation baseline questionnaires about the attitudes and beliefs of 21 715 women: 1445 randomly selected at the outset and 20 270 who were recalled for further tests . Thus, tParticipants completed sociodemographic and medical history details and four questionnaires: (1) one see  probing Logistic regression was used for binary outcomes and proportional odds logistic regression for ordinal outcomes. The proportional odds logistic model assumes that, for any dichotomy of the ordinal scale, all odds ratios of interest, comparing the odds of observing a ‘higher’ outcome between two groups, do not depend on the cut point of the dichotomy. The multiple explanatory variables included in these models were age group, education level, centre, anxiety level, poor psychological health, partnership status, family history of OC , family history of breast cancer , personal history of breast cancer and having used oral contraception. Education level was categorised into three groups according to the highest formal qualifications specified: tertiary education , secondary education  and no formal qualifications specified. Anxiety level was categorised in three groups: low, average and high. Low and high anxiety were, respectively, defined as an STAI trait score lower or higher than one s.d. from the mean score using the sample mean (36.5) and s.d. (10.1). The GHQ score covariate was dichotomised with a ‘case’ of probable psychological morbidity defined as a GHQ score of 4 or more. The results are reported as odds ratios with 95% confidence intervals (CI). Reported results for all regression models are adjusted for centre. Analyses restricted to the random sample were undertaken, but no qualitative differences arose.With such a large sample size, tests of the simple null hypothesis of no effect because of an explanatory variable led to many significant results. However, many of the estimated odds ratios, significant at the 5% level are close to 1 and too small to be of practical importance  of the 21 358 respondents gave the most accurate answer (1 in 70), whereas 50.1% underestimated the risk (1 in 500) and 7.6% overestimated the risk (1 in 12). When asked to estimate their own lifetime risk of developing OC, 38.9% (95% CI: 38.2–39.5) selected 1 in 100, 54.2% of 21 332 selected 1 in 500 and 6.9% selected 1 in 10. Estimation of a higher level of personal risk was associated with a personal history of breast cancer and a family history of OC and higher anxiety levels , and this was considerably more likely among respondents with a personal or family history of breast cancer and also associated with a higher level of education (The majority (84.4% 95% CI: 83.9–84.9) of 21 552 respondents were aware that most ovarian lumps turn out to be cysts rather than cancer and 13.8% did not know. Those who answered correctly were more likely to be younger, more educated and less anxious (Only 8.8% (95% CI: 8.4–9.2) of 21 205 respondents were aware that the chances of an ovarian lump being cancer are highest in women aged over 65 years. Correct answers were mainly associated with having a family history of OC, older age and tertiary education  had spoken to other members of their family about the risk of OC and this was more likely among those with a family history of OC than those without a family history . Participants were asked what they perceived to be the costs and benefits of screening and for their views on their susceptibility to OC . The majJust over two-fifths of the UKCTOCS participants had an accurate awareness of lifetime population risk of OC, but half underestimated the risk. As far as personal risk was concerned, higher levels of anxiety, a personal history of breast cancer or a family history of OC were associated with a higher estimation.With the exception of a family history of OC, there was a lack of awareness regarding specific risk factors and there were some common misconceptions. Only 26.7% knew of the increased risk of OC associated with a family history of breast cancer. Only 38.2% identified the increased likelihood of developing OC after the menopause  and <1 in 10 were aware that the chances of an ovarian mass being cancer are highest in women older than 65 years. Few recognised the association between lower parity and increased OC risk or the protective effects of breastfeeding. Even more importantly, there were mistaken beliefs; over a third associated having an abnormal cervical smear with an increased likelihood of developing OC and over a quarter of the participants believed taking the contraceptive pill increases the risk of OC.Despite the absence of data yet showing conclusive benefit of OC screening, most participants believed that a high uptake of it would reduce mortality and that screen-detected cancers would have an improved prognosis. The recent campaigns and publicity to improve awareness about the symptoms of OC are clearly important as only 41.1% recognised that it might be possible to identify symptoms sooner than waiting for screening.Ultrasonography is an invasive procedure and more than half (51.9%) of the participants felt that gynaecological examinations were embarrassing, which might ultimately have an effect on willingness to attend and acceptability of this form of screening. Although approximately two-thirds were concerned about getting OC, it was encouraging that few (15.2%) said that coming for OC screening would cause them to worry unnecessarily.The women participating in UKCTOCS were volunteers from the 1.2 million invited and their risk perceptions may reflect this bias. Participants' general knowledge of OC may also have been raised due to pre-recruitment information. In addition, volunteers for prevention or screening trials tend to be healthier and may be better educated and of higher socioeconomic status than the general population . TherefoWhen considering screening participation women need to consider their own risk of developing OC, as there are potential costs as well as benefits. The questionnaire asked about numeric estimates of lifetime risk, but the interpretation of probability information, is difficult  and it mThe risk factors included in the UKCTOCS questionnaire were selected >10 years ago and other risk factors for OC have since been described including obesity, current use of hormone replacement therapy, infertility, perineal talc use and endometriosis. Recent campaigns to increase OC awareness may also have increased public understanding.The results from UKCTOCS and other international studies regarding the ability of screening to detect OC in time to provide treatment that might influence survival, the best screening method, its acceptability to patients and other costs and benefits are still awaited. If the data do support implementation, then information including the risk perceptions of individuals who agree to OC screening may be important to our understanding of factors that motivate attendance, especially as there are reported inequalities in the use of existing breast and cervical screening services (As all the UKCTOCS participants had volunteered for the study, it is not surprising that most believed in the value of OC screening, but the majority of participants also believed that an increase in uptake would reduce mortality. This finding is perhaps surprising as women had received a detailed information sheet on the pros and cons of screening and had attended an information session during recruitment, when they were informed that the primary aim of UKCTOCS is to examine whether ovarian screening can reduce mortality. The belief that screen-detected cancers would have a better prognosis is perhaps more understandable given that this is the underlying premise on which OC screening is based. These beliefs highlight the importance of presenting balanced information about the limitations as well as the benefits of screening, which has been highlighted in recent controversies about breast screening . HoweverThe 6-year UKCTOCS follow-up study will be able to examine whether an underestimation of population risk leads to poorer attendance for screening. As well as improving information about OC risk and screening, more research needs to be performed on ways of assisting comprehension and understanding.If the risk of OC is underestimated in the general population, campaigns should be encouraged to increase awareness. At the same time, it is important that this should be put in perspective and women should be aware that OC is only the fifth most common cancer among women in the UK accounting for 5% of all female cancers . SpecifiAlthough having more educational qualifications, a family history of cancer and to some extent age were associated with improved knowledge about specific aspects of OC, there was a low level of awareness in general. Thus, the findings from this large study of UK women show the urgent need for improved public education about all aspects of OC. Opportunistic discussions with women about OC when being seen by their primary care physicians and information leaflets in appropriate settings may all contribute. These results provide important information for the development of awareness initiatives that might be needed for OC prevention and for screening programmes, if they are adopted.


MTHFR) gene has been hypothesized to increase breast cancer risk. However, results have been inconsistent, and few studies have reported the association by menopausal status or by intakes of nutrients participating in one-carbon metabolism. Our aims were to investigate whether MTHFR C677T was associated with postmenopausal breast cancer risk and whether this relation was modified by intakes of folate, methionine, vitamins B2, B6, and B12, and alcohol.The C677T polymorphism of the methylenetetrahydrofolate reductase  and 95% confidence intervals (CI) were calculated using unconditional logistic regression.We studied 318 incident breast cancer cases and 647 age- and race-matched controls participating in a nested case-control study of postmenopausal women within the VITamins And Lifestyle (VITAL) cohort. Genotyping was conducted for P for trend = 0.04). Evidence of effect-modification by intakes of some B vitamins was observed. The most pronounced MTHFR-breast cancer risks were observed among women with the lowest intakes of dietary folate (P for interaction = 0.02) and total  vitamin B6 (P for interaction = 0.01), with no significant increased risks among women with higher intakes.We observed a 62% increased risk of breast cancer among postmenopausal women with the TT genotype . Women with a higher number of variant T alleles had higher risk of breast cancer (MTHFR 677TT genotype is associated with a moderate increase in risk of postmenopausal breast cancer and that this risk may be attenuated with high intakes of some one-carbon associated nutrients.This study provides support that the  MTHFR gene is a common polymorphism (C677T) and is correlated with increased thermolability and reduced MTHFR activity [Methylenetetrahydrofolate reductase (MTHFR) catalyzes the irreversible reduction of 5,10-methylenetetrahydrofolate to 5-methyltetrahydrofolate, the primary circulating form of folate and methyl donor in DNA methylation. MTHFR is a critical enzyme in one-carbon metabolism, redirecting the pool of folate from DNA synthesis/repair to methylation. It is of interest because aberrations in DNA synthesis, repair, and methylation, have been implicated with cancer risk. The substitution of cytosine (C) with thymine (T) at nucleotide 677 in the activity .MTHFR C677T polymorphism has been largely inconsistent [The C677T polymorphism has been studied extensively, yet for breast cancer risk, three recent meta-analyses suggest that the association with the nsistent -5. Severnsistent ,4. Howevnsistent -5. A furMTHFR C677T polymorphism and breast cancer risk among postmenopausal women in a nested case-control study within the VITamins And Lifestyle (VITAL) cohort. We also examined whether this relation was modified by prediagnostic intakes of nutrients involved in one-carbon metabolism . Previously, in the VITAL cohort, we observed a protective association between folate intakes and breast cancer risk [We, therefore, investigated the relationship between the The VITAL cohort was principally designed to investigate supplement use and cancer risk. Details have been previously reported . BrieflyMTHFR. After these criteria, 318 postmenopausal VITAL women who had been diagnosed with incident breast cancer (invasive and in situ) remained.Breast cancer cases were identified by linkage to the SEER cancer registry. From baseline to December 30, 2003, 514 breast cancer patients were identified. To form the nested case-control dataset, we excluded women who reported a history of breast cancer (n = 48), had rare breast histologies , did not provide buccal cell samples needed for genotyping (n = 127), or did not complete the breast-cancer risk-factor page of the baseline questionnaire (n = 1), leaving 334 cases. We further excluded 12 women who were not postmenopausal and four women who failed to be genotyped for MTHFR, leaving 647 postmenopausal controls. This study was approved by the Fred Hutchinson Cancer Research Center Institutional Review Board (IRB), Seattle, Washington. Voluntary return of the questionnaire was considered implied consent.Controls were women who were not diagnosed with any type of cancer since baseline (based on linkage to SEER) and had not reported a history of breast cancer on the baseline questionnaire. There were 36,096 VITAL women without a cancer diagnosis (any type of cancer) since baseline and had never been diagnosed with breast cancer. Of these possible controls, 11,798 were excluded because they did not provide a buccal cell sample, and 185 were excluded because the breast-cancer risk-factor page of the baseline questionnaire was not completed. Two controls for each case were randomly selected from the remaining 24,113 possible  controls by frequency matching on age at baseline  and race resulting in 668 controls. We also ensured that the follow-up times of the controls  were greater than or equal to the follow-up times of the cases (time from baseline to breast cancer diagnosis). Information on deaths and moves out of the area were obtained by linking the VITAL cohort to the Washington State death files and the National Change of Address system. For this current analysis, we excluded 19 controls who were not postmenopausal and two controls whose samples failed to be genotyped for MTHFR C677T was determined using a microsphere-based, allele-specific primer extension (ASPE) assay followed by analysis on the Luminex 100 flow cytometer  as previously described [DNA was obtained from buccal cells collected from cytobrushes. Women who completed the VITAL baseline questionnaire were mailed a DNA kit containing three sterile cytobrushes, detailed instructions with pictures for use of kit, and a consent form. Genotyping was done by InterGenetics Incorporated . Genotyping of escribed . DNA wasInformation on diet, medical history, reproductive history, lifestyle factors, personal characteristics, and family history of breast cancer was collected at baseline using a 24-page self-administered questionnaire. Women were considered postmenopausal if they had a natural menopause with no periods in the year before baseline, had ever used postmenopausal hormones (PMH), had bilateral oophorectomy, or were 60 years or older at baseline. Because women with a hysterectomy without bilateral oophorectomy cannot report on menopause, they were considered to be postmenopausal if they had ever used hormone therapy or were 55 years or older at baseline.2, B6, and B12, multivitamins and alcohol reported on the baseline questionnaire were examined as potential effect modifiers of the MTHFR-breast cancer relationship. Intakes from diet over the past year were determined from a semi-quantitative food frequency questionnaire (FFQ), adapted from the Women's Health Initiative and other studies [stress/B-complex, antioxidant mixtures), including questions on years of use in the 10 years before baseline, days per week of use and dose per day. The validity and reliability of supplement reporting in this cohort has been previously examined [Intakes of folate, methionine, vitamins B studies -11. Part2 was not asked as an individual supplement, so only B2 as a multivitamin was considered, and therefore supplemental B2 was not considered alone. Methionine was determined from diet only. Average daily intake of dietary nutrients was calculated from the FFQ by multiplying the adjusted serving frequency  of each food/beverage item by its nutrient content and summing the nutrient contributions of all foods or beverages. The nutrient database used, the Minnesota Nutrient Data System for Research  [We analyzed B vitamins from diet, supplement and diet plus supplement  sources. Vitamin BMN, USA) , took inone-carbon micronutrient score by summing the z-scores of total folate, dietary methionine, and total vitamins B2, B6, B12 and then, dividing this score into high (upper median) and low (lower median) for categorical analyses. Women were excluded from the dietary and total (diet plus supplement) nutrient analyses if they did not complete all pages the FFQ or if their reported total energy intake was <600 or >4,000 kcal. Breast cancer risk factors and demographic variables were reported on the baseline questionnaire [Intakes of supplemental folate, other micronutrients, and multivitamins were averaged over 10 years, as previously described . We needP > 0.05). Odds ratios (OR) for breast cancer risk and 95% confidence intervals (CI) were calculated using unconditional logistic regression adjusting for the matching factors of age at baseline  and race . For analyses of the MTHFR-breast cancer association, we assigned the wild-type genotype MTHFR 677CC as the reference group. Linear trend was calculated by modeling the MTHFR genotype  as one term, ordinally.We compared baseline characteristics of cases and controls using Wilcoxon signed-rank tests (for continuous variables) or Chi-square tests (for dichotomous variables). Observed genotype frequencies were in Hardy-Weinberg equilibrium  were dichotomized by levels at or above the dose obtained by 10 year daily use of a standard (Centrum®) multivitamin versus below that level. Based on previous literature on breast cancer risk [MTHFR genotype and the dichotomous effect modifier and used a single reference group for the odds ratios. The group hypothesized to have the lowest risk was selected as the reference group. Tests for interaction were performed by entering the product term of the ordinal MTHFR variable and dichotomous effect modifier in a multivariable-adjusted model and using Wald statistic to obtain a P-value. All statistical analyses were performed using SAS, version 9.1, . P values < 0.05 were considered statistically significant and all statistical tests were two-sided.Intakes of folate, methionine, vitamins Bcer risk , and theMTHFR genotypes were: 677CC (46.5%), 677CT (43.9%), and 677TT (9.6%).Participants were on average 64 years of age and mostly Caucasian Table . As expeMTHFR 677TT genotype had significantly higher risk of breast cancer  than 677CC individuals (Table P for trend = 0.04), although there was no clear excess risk for the heterozygous MTHFR 677CT genotype   and when restricting to Caucasians  (data not shown).Postmenopausal women with the MTHFR-breast cancer association was modified by intakes of nutrients involved in one-carbon metabolism : 0.60 to 0.79). However, only two of these results showed a statistically significant interaction: the risk of breast cancer among women with the MTHFR 677TT genotype was significantly higher among women with low intakes of dietary folate (P for interaction = 0.02) and total B6 (P for interaction = 0.01) than those with higher intakes. Results for total vitamin B6 were particularly striking. Among women with high B6 intake, there was no MTHFR-breast cancer association. However, among women with low B6 intake, the TT genotype was associated with an approximate four-fold risk (OR = 4.03 = 4.47/1.11). Alcohol intake did not appear to modify the exposure-disease association (P for interaction = 0.22), but few VITAL women drank ≥ 10 g/d of alcohol.We next examined whether the sm Table . In gene6.In this moderate-sized, nested case-control study, we observed a 62% increase risk of breast cancer among postmenopausal women with the TT genotype. The most pronounced risks were observed among individuals with the TT genotype and lowest intakes of folate and vitamin BMTHFR C677T and breast cancer risk have been inconsistent. Our results are consistent with two [Results from 26 case-control studies ,16-40 inwith two ,19 of niwith two ,29,31 rewith two . Among 1with two ,29-31, twith two ,30. Two with two ,32 obserwith two ,36,37,39MTHFR C677T and nutrients, including alcohol [Eight studies, to-date, have examined interactions between  alcohol -23,25. A alcohol . However alcohol ,21,23.6 has an important role in one-carbon metabolism in that it acts as a cofactor for methionine synthesis and is a coenzyme of serine hydroxymethyltransferase, which is involved in nucleotide synthesis. We did not observe an overall statistically significant association between B6 intake and breast cancer risk, previously [Our results are biologically plausible. The TT genotype is associated with approximately 65% less activity than wild-type ,2. ReducMTHFR polymorphisms, such as A1298C and G1793A. However, data suggest that C677T is the major genetic determinant of MTHFR activity [Our study has some limitations. First, we did not have data on other activity . Second,MTHFR by multiple nutrients involved in the one-carbon metabolism pathway; data on supplement use also allowed for analysis by nutrient type . This study also adds to current knowledge regarding the association of MTHFR among postmenopausal women. An advantage of this study's prospective design is that it avoids recall bias for the recall of nutrients. An additional strength is that we had a broad range of information on possible confounders for adjustment. Also, selection bias is highly unlikely, because this was a prospective study and participants would not have known their future breast cancer status when deciding to give buccal cells or complete the breast cancer risk questions.Our study adds to the literature by being the first study, to our knowledge, examining effect modification of MTHFR 677TT genotype is associated with a moderately increased risk of postmenopausal breast cancer and with a substantial increase among women with low intakes of folate and vitamin B6. From a public health standpoint, these results are of interest in that they suggest that the increased risk associated with the TT genotype may be attenuated with intakes of some one-carbon associated nutrients.In summary, this study provides support that the ASPE: allele-specific primer extension; BMI: body mass index; CI: confidence interval; DNA: deoxyribonucleic acid; FFQ: food frequency questionnaire; IRB: Institutional Review Board; kcal: kilocalorie; MET: metabolic equivalent; MTHFR: methylenetetrahydrofolate reductase; PMH: postmenopausal hormones; OR: odds ratio; PCR: polymerase chain reaction; SD: standard deviation; SEER: Surveillance, Epidemiology, and End Results; VITAL: VITamins And Lifestyle.ERJ is a salaried employee of and holds stock options in InterGenetics Incorporated, but is not a major shareholder. A grant from InterGenetics Incorporated provided partial support for genotyping analyses. The other authors declare that they have no competing interests.SSM planned the data analysis, carried out all statistical analyses, interpreted the results, and drafted the manuscript. CMU contributed to the interpretation of the data and revised the manuscript for important intellectual content. ERJ directed the DNA isolation and the acquisition and finalization of the genotyping studies and revised the manuscript. EW conceived of the VITAL study, obtained funding, participated in the study's design and coordination, contributed to the interpretation of results, and revised the manuscript for important intellectual content. All authors read and approved the final manuscript.


Sir,et al is not adequately powered for this purpose.We would like to comment on the paper , that anet al's assertion that high risk population surveillance is ‘inefficient’, there remains a need for screening among younger women in this group, who understandably wish to delay prophylactic surgery in order to have children or to avoid premature menopause. This is evidenced by the fact that 60% of the women in the UK Familial Ovarian Cancer Screening Study (UKFOCSS) are aged under 50 years . We have recently obtained funding from Cancer Research UK and the Eve Appeal to screen a high-risk population of 3000 women for 5 years using 4 monthly rather than annual CA125 testing (phase 2 of UKFOCSS). The CA125 results will be analysed using the risk of ovarian cancer algorithm (ROCA) (Despite Oei Until UKFOCSS and the CGN study report in 2012, there is unlikely to be sufficiently reliable evidence to conclude that familial ovarian cancer screening is effective or ineffective. We therefore recommend that high-risk women in the US and UK unwilling to undergo prophylactic bilateral salpingo-oophorectomy take part in these studies if they wish to undergo screening.


Since the two components of adult height – leg length and trunk length – are poorly correlated with each other and appear to be influenced by different early life factors, examining their separate influence on breast cancer may provide additional insights into the mechanisms responsible for the positive association between adult height and breast cancer. In a cross-sectional study of 4286 women aged 60–79 years, in whom there were 170 cases of breast cancer, we found total height, leg length and trunk length were all modestly positively and linearly associated with breast cancer. The magnitudes of the associations of leg and trunk length were similar: fully adjusted odds ratio  of breast cancer for a one standard deviation (s.d.) increase in leg length 1.17  and for a 1 s.d. increase in trunk length 1.19 . Self-reported birth weight (available on 33% of the sample) was positively and linearly associated with breast cancer: fully adjusted odds ratio of breast cancer for a 1 s.d. increase in birth weight 1.30 . These associations were all independent of each other and other potential confounding factors and are likely to reflect different mechanisms by which factors operating prenatally and prepubertally influence breast cancer risk. Ecological studies of breast cancer have shown that geographic variations in incidence and mortality are associated with variations in population height . The majThe underlying mechanisms for the association between tall stature and breast cancer risk are not understood. The association may be explained by factors that jointly influence stature and breast cancer risk. These include hormones such as insulin-like-growth factor I (IGF-I)  women aged 60–79 years randomly selected from general practitioner lists of 23 British towns. Selection of towns, general practitioners (GPs) and participants was based on the methods used for the British Regional Heart Study of men . Leg length was taken as the standing height minus the trunk length. Weight was measured in light clothing without shoes to the nearest 0.1 kg using Soenhle portable scales. Waist circumference was taken as the midpoint between the lower rib and the iliac crest. Hip circumference was taken as the largest circumference below the waist.n=11). For women who had had an oophorectomy prior to their natural menopause, the date of the oophorectomy was taken as their date of menopause. For women who had had a hysterectomy without oophorectomy prior to their natural menopause, their age of menopause was assumed to be the median of the cohort – 50 years (n=11 breast cancer cases), and for women with breast cancer who did not provide details of their age at menopause, their age at menopause was assigned 50 years (n=3). The women were asked to report their birth weight in pounds (lb) and ounces (oz) in the self-completed questionnaire; they were not given the option of providing birth weight in prespecified categories. For comparisons with other studies, the self-reported birth weights were converted from lb to kg by multiplying by a factor of 0.4545.Three sources of data were used to determine breast cancer status: (i) women were asked if they had ever been told by a doctor that they had breast cancer and, if so, the date of diagnosis; (ii) diagnoses of breast cancer together with dates were extracted from the general practitioner medical records ; and (iii) all participants were flagged with the National Health Service Central Register (NHSCR), which provided details of cancer registrations. Anyone with a diagnosis of breast cancer from any one of these three sources was considered to be a prevalent case. Flagging of participants with the NHSCR was continuous, and is still ongoing, from the time of the baseline fieldwork (completed in March 2001). We included all cases from the NHSCR that were reported to us by the NHSCR up to November 2002  and that had been diagnosed prior to the date that the woman attended for baseline study examination. Cases were defined as pre- and postmenopausal based on self-report of age at menopause and the date of first diagnosis of breast cancer obtained from the GP record, cancer register or, if not available from these two sources, self-report (n=405), age at menopause (n=280), smoking (n=19) and osteoporosis (n=476). In all multiple logistic regression models, only those with complete data on all variables (n=3554) included in the fully adjusted model were included in each simpler model. In all analyses, robust standard errors taking into account possible nonindependence between women from the same town were used to estimate confidence intervals.Pearson's correlation coefficients were used to assess associations between total height and components of height. To illustrate the direction and shape of associations, age-adjusted means and prevalences of breast cancer and possible confounding factors were estimated across quarters of height and each component of height. These were estimated using multiple linear and logistic regression models with the age variable centred around the mean age value. Multiple logistic regression was used to assess the associations of the various anthropometric measures with breast cancer prevalence, with adjustment for potential confounding or intermediary variables. In these models, age, age at menarche, age at menopause, weight, waist-to-hip ratio and birth weight were entered as continuous variables. Osteoporosis , hysterectomy and/or oopherectomy  adult and childhood social class  and smoking  were entered as dummy variables. Of the 4286 women, 425 could not be assigned an adult social class and 545 could not be assigned to a childhood social class because they did not provide data on occupation. Although the participants were not specifically asked about unemployment, these women are likely to have been married to unemployed men (for those with missing adult data) and had fathers who were unemployed (for those with missing childhood data). This is consistent with the findings that women without these data on social class were more likely to smoke, more likely to be obese, were shorter and were more likely to have prevalent coronary heart disease than cohort members who provided these data , P=0.7). In total, 170 of the participants had a diagnosis of breast cancer from at least one source, giving an overall prevalence of 4.0% . The majority (87%) of these cases were identified from at least two sources. The age distributions of women with cancer identified by each source were similar – mean (standard deviation) age of women with breast cancer identified by self-report 68.5 (5.3), identified by medical record review 68.4 (5.3) and identified by cancer register 68.6 (5.3). Of the 170 cases, 39 (22.9%) were premenopausal and 131 (77.1%) were postmenopausal.The prevalence of breast cancer among women who were invited to take part in the study but did not respond was obtained from GP record reviews – the proportion of general practice recorded breast cancer did not differ between responders and non-responders  r=0.24 ). The weak correlation between the two components of height justifies considering their roles in the association between height and disease outcome separately.While both leg length and trunk length were strongly correlated with total height  and 0.76  respectively), leg length and trunk length were only weakly correlated with each other . After adjustment for age, smoking, weight, waist-to-hip ratio, age at menarche, age at menopause, osteoporosis, adult social class and childhood social class, a 1 s.d. increase in total height was associated with an increased odds of prevalent breast cancer: 1.25 . Both leg length and trunk length were associated with increased odds of prevalent breast cancer, with the magnitude of the associations being similar for both components of height: fully adjusted odds ratio  for 1 s.d. increase in leg length 1.17  and for 1 s.d. of trunk length 1.19 . Since age at menarche is associated with height and breast cancer, this could potentially act as an important confounder or explanatory factor. However, adjustment for menarcheal age alone did not importantly influence the associations between total height or any of the components of height and breast cancer. With additional mutual adjustment of leg length for trunk length and vice-versa both components of height remained independently associated with breast cancer prevalence: fully adjusted (including trunk length) odds ratio for breast cancer prevalence with 1 s.d. increase in leg length 1.14  and fully adjusted (including leg length) odds ratio for breast cancer prevalence with 1 s.d. increase in trunk length 1.16 . To assess possible selection bias arising through missing data, we estimated age-adjusted odds ratios for each anthropometric variable in the complete data set. The age-adjusted odds ratios for all women ; leg length: 1.14 ; and trunk length 1.19 ) were similar to those presented in When these associations were examined separately for premeno-pausal and postmenopausal cancers, there were no associations between height or each component of height and pre-menopausal cancer, but total height, leg length and trunk length were all positively associated with postmenopausal breast cancer , P=0.3). The mean (s.d.) of self-reported birth weights in this cohort was 3.28 (0.80) kg. Self-reported birth weight was weakly positively associated with both leg length and trunk length – age-adjusted Pearson's partial correlation coefficient for the association between birth weight and leg length 0.15  and between birth weight and trunk length 0.16 . In this subgroup of women with data on birth weights, birth weight was positively associated with breast cancer prevalence – age-adjusted odds ratio  of breast cancer for a 1 s.d. (0.80 kg) increase in birth weight was 1.33 . The association was only slightly attenuated in fully adjusted models: 1.30 . Additional adjustment for height had little effect: 1.29 .Of the 4286 participants, 1394 (33%) provided details of their birth weight. There were no differences in prevalent breast cancer between women who provided details of their birth weight and those who did not  Adjustment for birth weight did not significantly alter the associations between total height and components of height with breast cancer , the results were unchanged. Since socioeconomic position is strongly associated with adult height, we repeated the multivariable analyses with different indicators of socioeconomic position in childhood (car access and bedroom sharing) and adulthood (car access and housing tenure) and found no difference to the multivariable results presented in In this study, adult height, leg length, trunk length and birth weight were all modestly and positively associated with breast cancer. The magnitude of the association between total height and breast cancer was similar to that found in a number of prospective cohort studies (vs 55% older adults in the 1991 census). Further, no indicator of childhood or adulthood social class were found to confound our results. Response bias is, therefore, unlikely to have had an important effect on our results.Our response rate (60%) is moderate but the prevalences of GP recorded cases of breast cancer were similar between responders and nonresponders in our study. Height is significantly affected by socioeconomic position, but the social class distribution in our study sample is similar to that found for the 1991 census  and breast size (as indicated by cup size) have been found to predict breast cancer (The association between birth weight and breast cancer risk most likely reflects intrauterine exposures to maternal hormones that influence both foetal growth and intrauterine development of the mammary gland. Oestrogen would be one plausible hormone since exposures to high levels of intrauterine oestrogen lead to increased foetal size  and indiThe associations between different measures of foetal and childhood growth most likely reflect differing intrauterine and childhood environmental risk factors for breast cancer. These findings highlight the importance of foetal and early life exposures in the aetiology of breast cancer. Further research should concentrate on testing specific hypotheses concerning these associations as outlined in


Children whose twins have had leukaemia have a higher risk of contracting acute lymphoblastic leukaemia (ALL), confirming a prenatal origin of the disease. This association was not true when considering other types of affected first-degree relatives. Children whose fathers were diagnosed with testicular cancer have a higher risk of ALL. Although acute lymphoblastic leukaemia (ALL) is the most common neoplasia among children worldwide . PossiblThis population-based cohort study aims to investigate the risk of childhood ALL associated with a family history of leukaemia and other neoplasms in parents and/or siblings.Data from the Swedish Family-Cancer Database were used. This Database was described in detail previously , diagnosed between 1961 and 2002, and aged less than 20 years old were considered.Standardised incidence ratios (SIRs) for childhood ALL were calculated dividing the observed number of ALL by the expected number of cases. The SIRs were stratified by follow-up period , and age using 5 years categories. 95% Confidence intervals (CI) were obtained assuming a Poisson distribution .In the total population of 6 994 345 individuals, 1925 cases of childhood ALL were ascertained. Only 19 of those had a first-degree relative diagnosed with leukaemia, with a SIR for childhood ALL of 2.21 (95% CI: 1.33–3.45) . HoweverWhen investigating the risk of childhood ALL associated with histories of other types of cancer among first-degree relatives, only children with parents diagnosed with testicular cancer showed an increase in risk. The SIR of childhood ALL for such offspring was 3.12 (95% CI: 1.50–5.75) . Only thChildren with twins diagnosed with leukaemia had a much higher risk of contracting acute lymphoblastic leukaemia than the general population. This increase in risk was not observed with other type of affected first-degree relatives, such as parents or siblings (excluding twins).In this large study, the data were obtained through Swedish national registries. These registries were reported to be close to 100% complete , includiPrevious publications using information from the same Database reported no increase in childhood ALL when parents were diagnosed with leukaemia . A higheThis study adds substantially to previous studies, as it investigates the familial cancer risk of childhood ALL considering different types of leukaemia in affected first-degree relatives. Indeed, epidemiological studies have mainly considered family histories of other malignancies than ALL in first-degree relatives, such as solid tumours . In thosIn this article, a very high risk of childhood ALL was found in people with twins diagnosed with leukaemia. All the twin pairs for whom at least one of them was affected were females, suggesting that they were monozygotic twins. Commonly, increases in cancer risk seen in people with close relatives diagnosed with cancer are suggested to be associated with shared genes and/or post-natal environment. However, it has been mentioned that due to the young age at onset of childhood ALL and the latency expected for cancer evolution, it is possible that the disease originates before birth . It has In the present study, when considering 18 different cancer sites diagnosed in first-degree relatives, an increase of childhood ALL was found only in participants with parents diagnosed with testicular cancer. Due to the large number of cancers considered, it is possible that the higher risk observed in these children is due to chance. An unproven explanation could be that the treatment of fathers for testicular cancer with cisplatin caused permanent damage to sperm which could contribute to the higher risk of ALL observed in offspring . When coMore studies are needed to investigate whether the increased risk of ALL observed in children with twins diagnosed with leukaemia is associated with common familial factors such as shared genes and/or postnatal environment, or if it is more explained by prenatal factors.This population-based cohort study indicates that children with first-degree relatives diagnosed with leukaemia have a higher risk of contracting ALL. However, this increase in risk seems to exist only when the affected relative is a twin, confirming a prenatal origin of childhood ALL.


Radiotherapy for breast cancer reduces disease recurrence and breast cancer mortality. However, it has also been associated with increased second cancer risks in exposed sites.We evaluated long-term second cancer risks among 182 057 5-year survivors of locoregional invasive breast cancer diagnosed between 1973 and 2000 and reported to US NCI-SEER Program cancer registries. Multivariate Poisson regression was used to estimate the relative risk (RR) and excess cases of second cancer in women who had surgery and radiotherapy, compared with those who had surgery alone. Second cancer sites were grouped according to doses received from typical tangential breast fields.By the end of 2005 (median follow-up=13.0 years), 15 498 second solid cancers had occurred, including 6491 contralateral breast cancers. The RRs for radiotherapy were 1.45 =1.33–1.58) for high-dose second cancer sites  and 1.09 (1.04–1.15) for contralateral breast cancer (≈1 Gy). These risks decreased with increasing age and year of treatment. There was no evidence of elevated risks for sites receiving medium ) or low doses ). The estimated excess cases of cancer in women treated with radiotherapy were as follows: 176 (95% CI=69–284) contralateral breast cancers or 5% (2–8%) of the total in all 1+year survivors, and 292 (222–362) other solid cancers or 6% (4–7%) of the total.Most second solid cancers in breast cancer survivors are not related to radiotherapy. Substantial improvements in breast cancer treatment and screening signify that nearly 90% of women diagnosed with breast cancer survive for >5 years . As there is at least a 5-year lag period between radiation exposure and solid cancer induction (n=33 214) and restricted follow-up to attained ages <80 years. We also excluded women if it was unknown whether they had received radiotherapy (n=352), chemotherapy (n=1273) or hormonal therapy (n=1304); women who did not have breast cancer surgery or if surgery was unknown (n=4848) and women who had distant metastases at diagnosis or unknown disease stage (n=7853). As a number of women were missing more than one piece of information, the final cohort included 182 057 women.The cohort was composed of women who were diagnosed with a first primary invasive locoregional breast cancer reported to one of the nine SEER registries between 1 January 1973 and 31 December 2000  or breast-conserving surgery . Before 1983, the type of surgery performed was not available in SEER, but as breast-conserving surgery was rare, surgery before this date was assumed to have been mastectomy. The effects of chemotherapy and hormonal therapy were not assessed directly because of concerns regarding underascertainment in SEER , medium (0.5–0.99 Gy) and low (<0.5 Gy) dose. This dosimetry was conducted using thermoluminscent dosimeters in tissue equivalent phantoms assuming a standard treatment protocol with 50 Gy tumour dose and beam energy of 6 MV photons . Standardised incidence ratios (SIRs) were calculated by dividing the observed number of cancers (O) by the expected number (E). Tests of statistical significance were two sided and were based on Poisson exact methods (The observed (O) number of second cancers for each site was compared with the expected (E) number in women in the general population. The expected number of second cancers was estimated by multiplying age-, sex-, race- and calendar-period-specific cancer incidence rates in the nine SEER registries by stratum-specific person-years at risk and then summing across strata  and surgery alone (r=0) groups, then the statistical expectation of Or is given by θ and β are parameters to be estimated and 1+β is the RR associated with radiotherapy. The use of the expected number of cancers as an offset indirectly adjusts for potential confounding by attained age and attained calendar period . As the majority of women who receive breast-conserving surgery are recommended to receive radiotherapy, the proportion of women who had this type of surgery without radiotherapy in SEER was relatively small (17%), and this subgroup is unlikely to be a representative subset of breast cancer patients. Therefore, to obtain stable risk estimates in subgroup analyses according to surgery type, we used a combined surgery-only comparison group that included both breast-conserving surgery and mastectomy patients. Tests for heterogeneity between surgery types took account of the lack of independence of the comparison group  is the most relevant summary statistic for all breast cancer survivors. The excess absolute risk (per 10 000 person-years) was estimated by dividing the excess by the associated number of person-years.vs 8%).Overall, 38% of women with breast cancer had surgery+radiotherapy and 62% were treated with surgery alone . RadiothThe women were followed up for an average of 13 years. During the follow-up period (1978–2005), a total of 15 498 women developed a second primary solid cancer, including 6491 contralateral breast cancers . Women tP-trend<0.001) and with increasing year of treatment (P-trend=0.01), but increased with increasing time since diagnosis (P-trend<0.001) (P>0.5).For sites in the high-dose group, the RR for surgery+radiotherapy, compared with surgery alone, decreased with increasing age at treatment (d<0.001) . For patvs 1.18 (1.02–1.35)). For patients diagnosed with breast cancer in 1983 or later, lung cancer risk was higher after mastectomy (RR=1.43 (95% CI=1.15–1.77)) than after breast-conserving surgery, but was still significantly elevated (RR=1.20 (1.07–1.36)). For oesophageal cancers, the RR was 1.99 (1.37–2.88), and there was no difference according to surgery type. Overall, the risk for soft tissue cancers was 2.52 (1.67–3.81), but it was especially highly elevated for angiosarcomas (RR=13.7 (4.0–95.6)). The majority of angiosarcomas diagnosed in women who had surgery+radiotherapy were located in the thorax (n=12 out of 16).We further examined the risks for specific sites within the high-dose group that had the largest number of cases: lung, oesophagus and soft tissue. The RR of lung cancer was significantly increased (RR=1.38 (95% CI=1.26–1.51)) and was higher for ipsilateral than for contralateral lung cancer (1.54 (1.36–1.75) P-trend=0.03, P-trend=0.02), but was not clearly associated with time since treatment, stage or type of surgery.The RR for contralateral breast cancer in women who had surgery+radiotherapy, compared with those who had surgery alone, decreased with increasing age at first breast cancer treatment  excess cases of contralateral breast cancer in women who had surgery+radiotherapy, or 5% (95%CI=2–7%) of the contralateral breast cancers diagnosed in 1 year survivors . The excIn this large long-term study of breast cancer survivors, significantly increased RRs of second solid cancers after radiotherapy were observed for the group of sites that typically receive the highest radiation exposure (⩾1 Gy). There was no overall excess risk for the group of sites that typically receive lower radiation exposures (<1 Gy). In total, it was estimated that about 5% (95% CI=2–7%) of contralateral breast cancers and 6% (3–8%) of other solid cancers occurring among 1+year survivors could be related to radiotherapy. Risks were generally lower for more recent treatment periods and higher for younger ages at treatment.To our knowledge, this is the first study to assess and quantify the long-term risk of all solid cancers after breast cancer radiotherapy in the United States. Our site-specific results were broadly similar to those from a pooled analysis of 63 randomised clinical trials of breast cancer radiotherapy . This is likely the result of several factors, including reduced volume of irradiated tissues such as the lung, since the introduction of computed tomography planning . In an eThree earlier studies with individual dosimetry have reported an increased risk of contralateral breast cancer after radiotherapy in the youngest patients (<age 45) , and breast cancer treatment and radiotherapy techniques have changed considerably over the past three decades (Overall, 5–6% of second solid cancers in irradiated women were estimated to be attributable to radiotherapy exposure. Among all breast cancer survivors, this figure was 3%. Our findings suggest that most second solid cancers after treatment for breast cancer are related to other risk factors such as lifestyle or genetic factors. When women and their physicians make treatment decisions, the risk of radiotherapy-related cancer needs to be placed in perspective and balanced with the known tumour control and mortality benefits achieved from treatment.


P-trend ⩽0.0002). Compared to women with a stable weight (±2 kg), the relative risk for women who gained 15–20 kg was 1.50  1.06–2.13). The pooled RR per weight gain increment of 5 kg was 1.08 (95% CI 1.04–1.12). Weight gain was not associated with breast cancer risk in current HRT users, although, overall, these women experienced a much higher risk of breast cancer compared with nonusers. Our findings suggest that large adult weight gain was a significant predictor of breast cancer in postmenopausal women not taking exogenous hormones.We examined prospectively the association between weight change during adulthood and breast cancer risk, using data on 1358 incident cases that developed during 5.8 years of follow-up among 40 429 premenopausal and 57 923 postmenopausal women from six European countries, taking part in the European prospective investigation into cancer and nutrition study. Multivariate Cox regression models were used to calculate hazard ratios according to weight change (kg), defined as the weight difference between age at enrolment and age 20 adjusted for other risk factors. Changes in weight were not associated with premenopausal breast cancer risk. In postmenopausal women, weight gain was positively associated with breast cancer risk only among noncurrent hormone replacement therapy (HRT) users ( A secondary aim of this study was to evaluate whether the association between weight change and risk of breast cancer is modified by (a) initial weight or BMI at age 20, and (b) current HRT use among postmenopausal women.The EPIC is a multi-centre prospective cohort study designed primarily to investigate the association between nutrition and cancer. The EPIC cohort consists of 23 subcohorts in 10 European countries: Denmark, France, Germany, Greece, Italy, The Netherlands, Norway, Spain, Sweden, and the United Kingdom, thereby allowing comparisons of lifestyle and food habits among regions with very different cancer rates. Food-related and lifestyle questionnaires were administered and anthropometric measurements obtained from all participants at the time of enrolment (1992–2000). The 366 521 eligible female participants were mostly aged 25–70 years and recruited from the general population residing in a given geographical area, such as a town or a province , Greece, Italy (Varese), Sweden , and the United Kingdom, with available data on recalled body weight at age 20 or 25 years, and measured or predicted  anthropometric characteristics at study enrolment. Women were classified according to menopausal status at enrolment based on an algorithm as described elsewhere  had been reported to the common database at the International Agency for Research on Cancer (IARC), Lyon, based on information on complete follow-up data up till December 2001 or December 2002 in most of the centres. Mortality data were coded according to the 10th Revision of the International Statistical Classification of Diseases, Injuries and Causes of Death (ICD-10), and cancer incidence data were coded according to ICD-O-2. This analysis included 1358 invasive  breast cancer cases, of which 264 occurred in women who were premenopausal at recruitment and 1094 in postmenopausal women.Incident breast cancer cases were identified through population cancer registries  or by active follow-up , depending on the follow-up systems in each of the participating countries. The active follow-up procedure used a combination of methods including health insurance records, cancer and pathology registries, and via contact with participants and their next-of-kin. Women were followed from study entry (1992–2000) until first breast cancer diagnosis, death, emigration, or end of the follow-up period. By April 2004, 5984 breast cancer cases , United Kingdom, Greece, Sweden  and Denmark, and for the age of 25 years in Germany (Potsdam). Weight and height at recruitment were measured to the nearest 0.1 kg and 0.1 or 0.5 cm, respectively, with subjects wearing no shoes.−2). Subjects with a BMI of ⩾25.0 to <30.0 kg m−2 were classified as overweight and those with a BMI of ⩾30.0 kg m−2 were classified as obese , leisure physical activity (continuous score), height (continuous), current OC use (no/yes), and current hormone use (no/yes). Current hormone use refers to the use of menopausal hormones at the time of recruitment as derived from the country-specific questionnaires or during interviews, and includes oestrogen alone and combined oestrogen/progestin preparations (referred to as HRT use).Information on reproductive, sociodemographic, and lifestyle characteristics was obtained from the standardised health questionnaire at study entry. Other known risk factors included in this analysis were: age at menarche , age at first pregnancy , education , smoking status , alcohol consumption as ethanol in g dayt0 defined as the subject's age at recruitment, and exit time t1 defined as the subject's age at breast cancer diagnosis or censoring date. All multivariate models were stratified (option ‘strata’ in SAS-PHREG procedure) by age at recruitment and by study centre to be less sensitive against violations of the proportional hazards assumption, and simultaneously adjusted for the following established or potential breast cancer risk factors: weight at age 20, height, leisure physical activity (as continuous variables), age at menarche, parity, age at first birth, education, smoking status, alcohol consumption, current OC use and current HRT use. An indicator category for missing responses for each covariate was created to minimise the loss of observations due to missing covariate data.Cox proportional hazards models were used to estimate adjusted hazard ratios (HR) and 95% confidence intervals (CI) of breast cancer incidence for each weight change category. Weight changes were defined as follows: −2 and +2 kg as the referent category (stable weight); more than −2 kg as weight loss; and more than +2 kg as weight gain. The latter was further divided into five weight gain categories . Age was used as the underlying (primary dependent) time variable in the counting process formulation, with entry time −2, and for postmenopausal women 21.3 kg m−2) and (b) current HRT use . To test for interaction, we used a metric score with median weight change within categories of weight change as the exposure variable. All tests of statistical significance were two-sided and P-values of <0.05 were considered statistically significant. The statistical analyses were performed with the use of the PHREG procedure in the Statistical Analysis System (SAS) software package, version 9 .Trend tests were calculated on the basis of category-based scores, assigning a score from 1 to 7 to an individual according to the weight change categories. For country-specific analyses, weight change was treated as categorical and continuous variable. Potential effect modifications of the weight change–breast cancer association were examined by (a) BMI at age 20  years and accrued a total of 573 102 person years. The median age at diagnosis was 49.0 years for premenopausal and 63.0 years for postmenopausal women. Baseline characteristics are presented in vs 12.1 kg). This large difference can be in part attributed to the different enrolment age. In all, 22% of premenopausal women had lost weight between age 20 (25) and baseline examination. The mean (s.d.) weight gain for the remaining women was 9.0±8.3 kg. At age 20, the prevalence of overweight (BMI 25–30) was 8.2% and the prevalence of obesity (BMI ⩾30) was 1.6%. At age of enrolment, these prevalences had increased to 21.0 and 8.9%, respectively.Among premenopausal women, long-term weight gain was not associated with breast cancer risk, irrespective of adjustment for potential risk factors . Weight Initial weight or BMI at age 20 was inversely, but nonsignificantly associated with risk of breast cancer after adjustment for relevant covariates (data not shown), and there was no evidence of statistical interaction between BMI at age 20 and weight change in relation to breast cancer risk.Similar to premenopausal women, for postmenopausal women, the heaviest women at age 20 were from the UK and the lightest were from Italy . On averP=0.002). Only after stratification by current HRT use, stronger positive associations between weight gain and breast cancer became apparent among women not currently using HRT at recruitment, and the interaction term between HRT use and weight change approached statistical significance (P=0.10) ; 1.0 (reference); 0.98 ; 1.01 ; 1.00 ; 1.12 ; 1.22 ; trend (P=0.10) . Compare−2) and weight change in relation to breast cancer risk irrespective of HRT use.Similar to premenopausal women, initial weight and BMI at age 20 were inversely, but nonsignificantly, associated with risk of postmenopausal breast cancer both before and after adjustment for weight change or other covariates (data not shown). There was no evidence of a significant interaction between BMI at age 20 , but with a limited follow-up period (5.8 years). Re-examination of our data after the removal of pre-existing disease by excluding individuals diagnosed during the first year of follow-up did not materially change the main findings, but did attenuate the association between weight loss and breast cancer risk among premenopausal women. Extended follow-up will allow us to evaluate the effect of excluding likely pre-existing disease with more confidence. Furthermore, it cannot be ruled out that some confounding bias is still present, due to the lack of inclusion of other unknown potential risk factors.Some limitations of these data should be noted. Our conclusions are based on results derived from a large cohort (cases vs weight gain over a lifetime, indicating no differential effect on breast cancer risk (We were unable to evaluate in more detail the effect of weight gain at different ages, other than that of weight at age 20 and weight at enrolment. To date, only a few studies have been able to disentangle the effects of weight at different ages In conclusion, these results show that large weight gain among postmenopausal women who do not use HRT is associated with an increased risk of breast cancer. Among current HRT users, who overall are at a higher risk of breast cancer than non-HRT users, weight gain is not associated with risk. These data provide further support for the positive association of adiposity with breast cancer among postmenopausal women. As obesity has reached epidemic proportions and is also a modifiable lifestyle factor, avoidance of weight gain throughout life is a salient measure for prevention of breast cancer.


In the latter group, the odds ratio for the highest vs the lowest quintile of enterolactone was 1.7 , 0.8–3.4; P-value for trend=0.05) and after adjustment for known risk factors for breast cancer was 1.6 . We observed a moderate positive correlation between serum enterolactone and serum sex hormone-binding globulin in postmenopausal women (r=0.29 in controls (P<0.001) and r=0.14 in cases (P=0.04)), but no correlation with oestrogens or androgens. These results do not support a protective role of circulating lignans, in the range of levels observed, in the development of breast cancer.It has been proposed that phyto-oestrogens protect against breast cancer. Lignans are the main class of phyto-oestrogens in Western diets. We conducted a case–control study of breast cancer and serum levels of the main human lignan, enterolactone, nested within a prospective cohort study, the New York University Women's Health Study. Serum samples collected at enrollment and stored at −80°C were used. Among 14 275 participants, 417 incident breast cancer cases were diagnosed a median of 5.1 years after enrollment. Cohort members individually matched to the cases on age, menopausal status at enrollment, serum storage duration and, if premenopausal, day of menstrual cycle were selected as controls. No difference in serum enterolactone was observed between postmenopausal cases (median, 14.3 nmol l It has been suggested that phyto-oestrogens protect against hormone-dependent cancers, in particular breast cancer, through their weak agonist/antagonist oestrogen properties as well as their actions on oestrogen synthesis and bioavailability . In addiA number of studies have focused on isoflavones and examined the association of intake of soy or soy protein with breast cancer risk, mostly in Asian populations (The association between circulating levels of the main lignan enterolactone and breast cancer risk was assessed in a case–control study nested within the New York University (NYU) Women's Health Study, a prospective cohort study of cancer, hormones, and environmental factors. The study focused exclusively on enterolactone based on the results of a preliminary study that examined the distribution and long-term reliability of serum measurements of the two main lignans, enterolactone and enterodiol, and of a number of isoflavonoid phyto-oestrogens. With the exception of enterolactone, phyto-oestrogen concentrations were below the detection limits for a large proportion of subjects, and the temporal reliability coefficients were low, indicating that our cohort was not well suited to study the effect of these compounds on disease risk , date of enrollment (±3 months), and if premenopausal, day of menstrual cycle at the time of blood donation. For the present study of enterolactone and breast cancer, one control was selected at random between the two controls  selected for the study of sex hormones in order to limit study costs. In addition, the following exclusion criteria were applied to both cases and controls: (1) use of antibiotics in the 4 weeks preceding blood donation because such use has been shown to almost completely eliminate the formation of enterolactone from plant precursors in the gut (μl of 3H-oestradiol glucuronide (enterolactone glucuronide is not available) was added to the tubes containing 200 μl of plasma and equilibrated for 30 min at room temperature. Then, 200 μl of acetate buffer 0.1 M pH 5.0 containing 2 U ml−1 sulphatase and 0.2 U ml−1β-glucuronidase were added. After mixing, the tubes were incubated overnight at 37°C. After hydrolysis, the free enterolactone was extracted twice with 1.5 ml of ethyl ether. The separation of the phases was carried out by freezing the water phase. After evaporation of the ether, 200 μl of the TR-FIA assay buffer was added and a 20 μl aliquot taken for calculation of recovery and two 20 μl aliquots used for the time resolved fluoroimmunoassay. Samples from a case and her matched control were always assayed in the same batch. The laboratory analyses were performed blind, and all of the batches were analysed with two quality control samples used throughout the whole procedure, and three samples used for the immunoassay step only. Two samples from a common pool that were labelled to preclude their identification were interspersed every 11 matched sets to assess the laboratory precision. The mean of the within-batch coefficient of variation for the seven batches was 8.9%.Enterolactone was measured by time-resolved fluoroimmunoassay  between case and control subjects, we used a mixed-effects regression model to take into account the matched design . P-valueP-values correspond to enterolactone variables treated as ordered categorical variables. Analyses were also performed on the continuous variables. Analyses were conducted using the SAS Release 8.02 and EGRET statistical packages. All P-values are two-sided.To compute odds ratios (ORs), serum measurements were categorised into quintiles separately for pre- and postmenopausal subjects, using the frequency distribution of the cases and the controls combined. The data were analysed using conditional logistic regression . Odds ran=24), diagnosis of coronary heart disease (n=3), and use of antibiotics in the 4 weeks prior to study enrollment (n=11). Six additional cases were excluded because the volume of serum was insufficient for the assay. As a result, 417 cases are included in this report. Nine participants became cases after being selected as controls and are included both as cases and controls in the appropriate matched sets . Although no longer statistically significant, this difference was observed in both pre- and postmenopausal women. Among women premenopausal at blood donation, cases tended to be older at first full-term pregnancy (P=0.08), reported more often a family history of breast cancer (P=0.001) and were less likely to have used hormone replacement therapy than controls (P=0.05). Among postmenopausal women, cases tended to have a higher weight (P=0.01) and body mass index (P=0.09) than controls.−1) than in controls . When levels were examined according to menopausal status, no difference was observed between postmenopausal cases and controls, whereas premenopausal cases had higher levels  than their matched controls .P=0.05). The OR was 1.7  in the highest quintile, which was slightly reduced to 1.6 (0.7–3.4) after adjusting for the following known risk factors: age at menarche, family history of breast cancer, parity, age at first full time pregnancy, height and body mass index (P for trend=0.13). Adjusting further for use of oral contraceptives and hormone replacement therapy did not materially affect the OR but reduced the number of matched sets available for analysis to 139  and are therefore not presented. Similar results were obtained when log-transformed enterolactone was used on the continuous scale. P<0.001) than in cases . This positive correlation persisted after adjusting for body mass index, a strong predictor of circulating SHBG: the partial correlation coefficients were 0.26 (P<0.001) in controls and 0.13 in cases (P=0.09).There was no evidence of correlation between enterolactone and oestrogens or androgens: all correlation coefficients were less than 0.10 , −1), and very high (>39.1 nmol l−1) concentrations  has been shown to inhibit both formation and growth of breast tumours in rats (Enterolactone at plasma levels that could easily be achieved in humans (400 nmol l in rats . In cell in rats . AdlercrPhyto-oestrogens could also affect the risk of breast cancer in premenopausal women through changes in the menstrual cycle length. Longer cycles have been shown to be associated with a reduced risk (In three of the prospective studies, an increase of breast cancer risk was observed at high levels of enterolactone, although the results were not significant in the Dutch study and only marginally significant in our study. These results suggest a need for further studies, both experimental and epidemiological, to better understand the overall impact of lignans in general, and enterolactone in particular, on the development of breast cancer. It also suggests caution against large dietary intake of flaxseed or use of flaxseed supplements for prevention: flaxseed is the oilseed with the richest amount of lignan precursors, and flaxseed supplementation can more than double serum enterolactone concentration (


Findings have been inconsistent on effects of adolescent body size and adult weight gain on risk of breast cancer in young women. These relations were examined in a population-based case control study of 1590 women less than 45 years of age newly diagnosed with breast cancer during 1990–1992 in three areas of the US and an age-matched control group of 1390 women. Height and weight were measured at interview and participants asked to recall information about earlier body size. Logistic regression was used to estimate the relative risk of breast cancer adjusted for other risk factors. Women who were either much heavier or lighter than average in adolescence or at age 20 were at reduced risk. Weight gain after age 20 resulted in reduced risk, but the effect was confined to early-stage and, more specifically, lower grade breast cancer. Neither the risk reduction nor the variation by breast cancer stage or grade was explained by the method of cancer detection or by prior mammography history. These findings suggest that relations between breast cancer risk in young women and body weight at different ages is complex and that the risk reduction with adult weight gain is confined to less aggressive cancers. © 1999 Cancer Research Campaign


To assess risk factors for cervical adenocarcinoma data were collected in a case-control study of 39 cases and 409 controls conducted in the greater Milan area. Questions were asked about personal characteristics and habits, gynaecologic and obstetric data, history of lifetime use of oral contraceptives and other female hormones, and general indicators of sexual habits . The relative risk of cervical adenocarcinoma increased with number of births and abortions, early age at first birth and early age at first intercourse. These estimates did not materially change after adjustment for the potential reciprocal confounding effect. Further, there was a positive association with overweight, but an apparent association with lower education was not significant. No relationship emerged with oral contraceptive use. Thus, despite the similarities with the epidemiology of squamous cell cancer, reproductive patterns and other factors related to the risk of endometrial cancer  seem to play an important role in the risk of adenocarcinoma of cervix uteri.


P = 0.29). The cancer screening rates and other variable values had fairly balanced distribution between the 2 groups. Having consanguineous parents decreases the risk of breast cancer especially in younger women, risk of cervical cancer being unaffected. © 2001 Cancer Research Campaign http://www.bjcancer.comMarriages between third-degree and more distant relatives are common in many parts of the world. Offspring of consanguineous parents have increased morbidity and mortality related to recessive gene disorders. In a population with a high frequency of consanguinity, we examined the frequency of breast cancer (related in part to tumour genes) and cervical cancers (related to virus infection) among offspring of consanguineous and non-consanguineous parents. Study was done prospectively in the United Arab Emirates. Selected were married female citizens, ages 40–65, who attended 12 primary health care clinics for whatever reason. In a face-to-face interview, subjects were asked: (a) about consanguineous marriages in family; (b) if they have or have had breast or cervical cancer; (c) about family history of cancer, cancer screening and other parameters. Tumour diagnosis was confirmed by review of medical records. Of 1750 women invited into study, 1445 (79%) could be used in analysis. Among 579 (40%) women of consanguineous and 866 (60%) of non-consanguineous parents there were 24 and 54 with breast cancer, respectively . In the 40 to 50 age group, breast cancer reported 13 of 446 women of consanguineous and 37 of 633 of non-consanguineous parents . Cervical cancer had 15 women in consanguineous and 32 in non-consanguineous group . Number of families with history of breast cancer in consanguineous and non-consanguineous group was 21 and 23, respectively (


Among these women, 251 subsequently developed cancer, of whom 96 developed breast cancer.After adjustment for age at entry, height, weight, age at menarche and first birth with nulliparity, there was a highly significant relationship between elevated GGT and breast cancer risk. In the highest quartile, the hazard ratio (HR) was 2.17 : 1.19, 3.93). When subdivided by menopausal status, there was a reduced non-significant effect in postmenopausal women, whereas for premenopausal women in the highest quartile, HR was 3.81 . Premenopausal women with serum GGT levels above the normal range had a significantly elevated HR of 4.90 .−1) are at increased risk of breast cancer and might benefit from close surveillance, possibly with breast magnetic resonance imaging scans. Serum GGT may mark previous exposure to carcinogens and lead to the identification of DNA adducts involved in mammary carcinogenesis.These results suggest that premenopausal women with high normal (above median) serum GGT or elevated levels (⩽40 IU l The enzyme is widely present on the external surface of most cells but has been particularly studied as a marker of hepatic dysfunction  for cancer incidence with increasing levels of GGT, so that among those with levels >72 U l−1, HR was 1.43 : 1.28–1.61). In terms of site specificity, elevated GGT was associated with cancers of the digestive tract, respiratory tract, breast/female genital organs and haematopoetic system.Recently, To investigate the relationship between GGT levels and risk of malignancy, particularly breast cancer, assays have been carried out on serum samples from 1888 women who participated in the Guernsey Breast Cancer Cohort Study, of whom 251 were subsequently diagnosed with cancer.The Guernsey Cohort Study has been underway since 1961 with the aim of identifying risk factors for breast cancer in a normal population . After centrifugation, aliquots of both serum and plasma were prepared. Aliquots of 2 ml were stored at −20°C and subsequently assayed in batches. Serum samples were used in this study. Serum GGT was measured by JPS Laboratories  using a routine kinetic colorimetric method. The coefficient of variation was 6% during the assay period.Assays were performed on 1888 samples that were randomly selected. However, 68 women had been previously diagnosed with cancer (excluding non-melanoma skin cancer), and a further three were found to have breast cancer during the study-screening process. These were excluded from the analysis. A further 14 samples were eliminated during the statistical analysis because of missing values in covariates. Thus, the final analysis was performed on 1803 samples.Follow-up of the cohort is still active and information on cancer incidence is obtained regularly through pathology reports from the only pathology laboratory in Guernsey, as well as from death certificates and data from the South West Cancer Registry. The study was approved by all relevant ethics committees and the women provided written consent.Cox proportional hazards regression models were fitted to estimate HRs for risk of any cancer (except non-melanoma skin cancer), risk of breast cancer and risk of any cancer excluding breast, using the Stata statistical package . For breThe distribution characteristics of GGT in this population are shown in The features of the 1803 participants, including the 96 subsequently diagnosed with breast cancer, are given in To determine whether GGT level was related to overall risk of cancer in this cohort, an analysis was conducted on all incident cancers. There were 251 cancers . Results are shown in As −1) were compared with those within the normal range, HR was 4.90  P=0.001. There was no relationship between disease interval and levels of GGT.The results were paradoxical. As This study found a significant relationship between GGT and overall risk of cancer, which disappeared when breast cancer cases were removed from the analysis. After adjustment for known risk factors, including weight, there was an independent significant relationship between elevated GGT and subsequent risk of breast cancer. However, no effect was seen in postmenopausal women, whereas for premenopausal women, HR was 3.82 for the highest quartile. Those premenopausal women with serum GGT levels above the normal range had a significantly elevated HR of 4.90.A limitation of the study is that we do not have information of alcohol intake among the participants. However, the blood samples were collected between 1986 and 1990 before there was a major increase in alcohol use in women.Because this was a prospective study, with samples being taken before development of breast cancer, it is likely that GGT elevation was not a result of malignant transformation, but possibly had a role in this process. Although GGT has been implicated in malignancy, this is the first study to show the specific effect in terms of breast cancer risk.γ-Glutamyl transferase catalyses hydrolysis and transpeptidation of extracellular GSH, thereby having a central role in maintaining glutathione homeostasis. GSH is an antioxidant, protecting cells against oxidative stress. Generation of the more reactive glycyl-cysteine reduces ferrous to ferric iron, thereby starting an iron redox cycling process (γ-glutamyl bond in GSH, areas of increased GGT may result in metal ion reduction and redox cycling. GSH/GGT-dependent pro-oxidant reaction has been found to modulate the transduction of proliferative/apoptotic signals (in vitro ( process . As GGT Studies on human breast cancers have given mixed results, probably because menopausal status and age were not taken into account. It is of particular interest that GGT is associated with increased risk when the woman is premenopausal. In addition, GGT did not seem to be associated with an increased risk for other cancers in this cohort. This raises the possibility of identifying those women with high normal or elevated serum levels. In the absence of evidence of alcohol abuse, these could be selected for closer surveillance. The method of surveillance is still uncertain. Routine clinical examination has not been shown to reduce mortality from breast cancer, nor has mammography in younger women. A recent Consensus Conference on high-risk women, mostly because of their family history, advised that breast magnetic resonance imaging should be performed annually but not before the age of 25 years (


P for trend=0.04). Our results indicate that exposure to female sex hormones does not protect against non-cardia stomach cancer and does not explain male predominance. The observed moderate inverse relationship between parity and cardia cancer may be mediated by non-hormonal factors and warrants further study.We investigated stomach cancer risk by anatomic sub-site in relation to parity, as a marker for higher exposure to sex hormones, in a case–control study, nested within a cohort of 2 406 439 Swedish women born in 1925 or later and followed from 1970 or age 30 until emigration, death, any cancer diagnosis, or through 2004, whichever occurred first. We identified 286 cardia and 2498 non-cardia stomach cancer cases with five matched controls for each case. Cross-linkage with the Multi-Generation Register provided information about reproductive history. Using conditional logistic regression models for estimating odds ratios (ORs) and corresponding 95% confidence intervals (CIs), adjusted for education level and occupation, we found no association between any aspect of parity and non-cardia stomach cancer . However, a 30% risk reduction for postmenopausal cardia cancer  was noted among parous relative to nulliparous women and the risk for premenopausal cardia cancer fell with increasing number of children ( Helicobacter pylori infection and tobacco smoking  were introduced, or later, constituted our cohort. NRN, a unique identifier assigned to all Swedish residents at birth or upon immigration, was used for unambiguous linkages with several nationwide registers. The Historic Population, Migration, and Causes of Death Registers were used to obtain information about marital status and dates of emigration and death, respectively. The Swedish Cancer Register, established in 1958 and more than 98% complete  with their corresponding 95% confidence intervals (CIs) as the measure of association between parity or age at first birth and risk. We first estimated crude ORs (inherently adjusted for year of birth) and then multivariately with additional adjustment for highest attained education , university education (>12 years)) and occupational class . We also considered place of residence  as a possible confounder, but as this did not materially affect our estimates, it was not included in the final models.We further studied the effect of number of children . Age at first birth was categorised based on the quartile values derived from the entire cohort . Dose–response relationships were studied only among parous women, and multivariate models mutually adjusted for parity and age at first birth. We used median values of age at first birth for each stratum and also parity as a continuous variable for trend analyses. Women were categorised as pre- or postmenopausal by using age 50 as the cut-off point, based on the approximate median value in several studies  and 1961 may be missing. As this could lead to some misclassification, we performed an additional analysis restricted to women born in 1946 or later and thus unlikely to have given birth to children before 1961.It is well known that unmarried status and living as a single are associated with poorer general health among both men and women, including higher risks for lifestyle-related cancers such as stomach cancer. We therefore reanalyzed data restricted to those who were married at cancer diagnosis  to check if our results were affected by marital status.P-value of the interaction term in the conditional logistic regression model was used as a test of homogeneity to compare ORs between pre- and postmenopausal women. All statistical analyses were performed using the SAS software version 9.1 . This study was approved by the Regional Ethics Committee at the Karolinska Institutet.The We identified 2498 women with non-cardia stomach and 286 women with cardia cancer between 1970 and 2004, and selected 12 490 and 1430 matched controls, respectively. The mean age of the non-cardia cases was slightly lower (57.0) than that of cardia cancer cases (59.4) .P-value of the test of homogeneity of results of pre- and postmenopausal cancers was 0.02. Analyses by parity or age at first pregnancy among parous women did not show any significant variation in ORs. Restriction to married women or to those born in 1946 or later produced no important departure from overall results.Parous women had the same risk for non-cardia stomach cancer as nulliparous women  . HoweverP=0.04) also for premenopausal cardia cancer . No conspicuous variation was observed in relation to age at first pregnancy. Restriction to cases and controls who were married at cancer diagnosis revealed stronger association , whereas restriction to women born in 1946 or later produced no noteworthy change, albeit based on very small numbers (data not shown).We found a 30% decreased risk of cardia cancer among parous compared with nulliparous women  . This inSocioeconomic status in our investigation was rather crude, since it combined two different classifications used in, respectively, the 1970 and 1990 censuses. We also repeated the analyses using only 1990 census information because this classification was finer . The point estimates thus obtained were almost identical to those in the main analyses (data not shown). However, as a considerable number of cases and controls who emigrated or died before 1990 were excluded in these supplementary analyses, we only present the results of analyses using the crude socioeconomic classification.This nested case–control study gave no indication that pregnancies confer protection against non-cardia stomach cancer, whereas some support for an inverse association between parity and cardia cancer risk emerged.The large sample size and the population-based design constitute the strengths of this study. Case ascertainment through the nationwide Swedish Cancer Register was virtually complete. There was possible misclassification of exposure due to missed offsprings who died before 1961, but additional analyses restricted to women born in 1946 or later suggested that this is unlikely to have importantly affected the results. However, abortions, miscarriages, and stillbirths were not recorded. This non-differential misclassification may have slightly shifted ORs towards the null value. However, small numbers limit the interpretation of our results for cardia cancer.An important limitation, as in most register-based studies, is the lack of information on some potential confounding factors such as smoking, a risk factor for stomach cancer, and it is conceivable that mothers smoke less than childless women. However, available data on smoking from other large cohorts of Swedish women do not support this  – not necessarily hormonal – that affect cancer development in the cardia and in the rest of the stomach differently. The hypothesis, therefore, that oestrogens play a role in stomach cancer aetiology is given no real support. On the other hand, parity may be an imperfect marker of the sex hormone exposure that is potentially relevant; women are continuously exposed to oestrogen during their fertile life, for up to 40 years, and the pregnancy boosts, although impressive in terms of dose, may be too short-lived or inadequately timed for an appreciable effect on gastric carcinogenesis.In conclusion, our results point fairly persuasively against a significant inhibitory role of oestrogen in non-cardia stomach cancer. Previous evidence of an inverse association with stomach cancer risk was, at best, only suggestive and with our study weighs towards no association. However, our finding of a significantly decreased risk of cardia cancer among parous, particularly postmenopausal women, warrants further study.


To our knowledge, these are the first prospective findings linking the intake of marine n-3 fatty acids to breast cancer protection.We investigated the effects of individual fatty acids on breast cancer in a prospective study of 35 298 Singapore Chinese women aged 45–74 years, who were enrolled during April 1993 to December 1998 . At recruitment, each study subject was administered, in-person, a validated, semiquantitative food frequency questionnaire consisting of 165 food and beverage items. As of December 31, 2000, 314 incident cases of breast cancer had occurred. We used the Cox regression methods to examine individual fatty acids in relation to breast cancer risk, with adjustment for age at baseline interview, year of interview, dialect group, level of education, daily alcohol drinking, number of live births, age when menstrual periods became regular, and family history of breast cancer. Consumption of saturated, monounsaturated or polyunsaturated fat overall was unrelated to risk. On the other hand, high levels of dietary n-3 fatty acids from fish/shellfish (marine n-3 fatty acids) were significantly associated with reduced risk. Relative to the lowest quartile of intake, individuals in the higher three quartiles exhibited a 26% reduction in risk (relative risk (RR)=0.74, 95% confidence interval (CI)=0.58, 0.94)); RRs were similar across the top three quartiles of intake . Overall, there was no association between n-6 fatty acids and breast cancer risk. However, among subjects who consumed low levels of marine n-3 fatty acids (lowest quartile of intake), a statistically significant increase in risk was observed in individuals belonging to the highest  The overall effect of dietary fat on breast cancer risk has been much studied with inconclusive results . On the Relatively sparse data are available on the potentially opposing and interactive effects of n-3 and n-6 fatty acids on breast cancer in humans (n=35 298) were women. The 564 women who reported a history of cancer at baseline were excluded from the present study. The study was approved by the institutional review boards of the University of Southern California and the National University of Singapore.The Singapore Chinese Health Study has been described previously , fresh shellfish , dried/salted fish , and canned fish . The average portion weight (without bone) for fresh fish was approximately 60 g and for fresh shellfish, dried/salted fish, and canned fish approximately 35, 10 g, and 60 g, respectively.in situ or stage I) and 188 advanced . In 31 cases, the stage was unknown due to lack of access to medical charts. In all, 93 patients were premenopausal at baseline, the remaining 221 postmenopausal.Incident cancer cases and deaths among cohort members were identified through linkage of cohort files with databases of the nationwide Singapore Cancer Registry . Levels of various dietary fats in individual subjects were computed from fat contents of food items listed in the Singapore Food Composition Table  and docosahexaenoic acid (DHA)) P-values. Study subjects were grouped into quartiles based on the distribution of the entire female cohort  in multivariate models involving postmenopausal women only did not appreciably alter the fish– (marine n-3 fatty acids) breast cancer association. Therefore, all RRs related to postmenopausal women in this report were unadjusted for BMI.In all analyses, we adjusted for the following potential confounders: age at baseline interview (years), dialect group (Hokkiens or Cantonese), year of interview (1993–1998), level of education , daily alcohol drinker , family history of breast cancer , number of live births , and age when period became regular . Inclusion of body mass index (BMI)  and Epilog windows version 1.0  statistical software packages. All Factors positively associated with risk were level of education, BMI , age at first birth, and family history of breast cancer. Factors inversely associated with risk were age when period became regular and number of live births  in this study reported consuming some seafood. On average, the mean fish/shellfish intake is 52 g/day. When women were grouped by intake of marine n-3 fatty acids in quartiles, those in the highest consumed, on average, 80 g of fish/shellfish per day, while those in the lowest had 25 g. Consumption of marine n-3 fatty acids was positively associated with the percentage of total energy from fat, saturated, monounsaturated, and polyunsaturated fatty acids, but inversely associated with alcohol consumption  .In seven of the 314 cases, there was a family history of breast cancer and five of these belonged to the lowest quartile of intake of marine n-3 fatty acids, yielding an RR of 4.2 (95% CI=1.7–10.4) for a family history of breast cancer in this dietary subgroup. The corresponding RR (95% CI) among those in the higher three quartiles of intake of marine n-3 fatty acids was 0.7 (0.2–2.9).In order to explore the possibility that the observed inverse association between fish intake and breast cancer might be an artefact due to the decreased consumption of fish in subclinical cancer patients, we repeated all analyses after excluding all cancer occurrences and person-year counts during the first 2 years of follow-up. Results were comparable to those based on a complete follow-up of the entire cohort (data not shown).In this study population, fish and shellfish account for roughly 30% of dietary n-3 fatty acid consumption. Other major sources of n-3 fatty acids are legumes (10%), grain products (22%), and cooking oils (10%). The major sources of n-6 fatty acids are meats (10%), grain products (20%), and cooking oils (40%).The present study found no evidence of an association between breast cancer and total, saturated, monounsaturated, or polyunsaturated dietary fat intake, in agreement with previous studies, including a meta-analysis of seven epidemiologic cohort studies from four countries totalling more than 330 000 women and nearly 5000 incident cases of breast cancer, as well as more recent analysis . No association was found in the four studies from the US (P for trend=0.05)  can reduce breast cancer risk by 25%. Finally, consistent with experimental results, our data suggest that high consumption of n-6 fatty acids was associated with increased risk of breast cancer among women consuming low levels of marine n-3 fatty acids.