We investigated whether dietary intakes of total fat, monounsaturated fat (MUFA), polyunsaturated fat (PUFA) and saturated fat (SFA) were associated with breast cancer risk in a prospective cohort of 49 261 Swedish women (30–49 years at enrolment), which yielded 974 breast cancer cases by December 2005. Further, we evaluated if associations differed by oestrogen and/or progesterone receptor tumour status. Total fat, MUFA, PUFA or SFA were not associated with risk overall. However, women in the highest MUFA and PUFA quintile intake had a reduced breast cancer risk after age 50 years  compared to women in the lowest quintile. The associations did not differ by oestrogen or progesterone receptor status. Despite the negative findings, type of fat during premenopausal years may have later differential effects on risk. Animal, ecological and some case–control studies suggest that a high consumption of total dietary fat increases breast cancer risk . There i+/PR+ but negatively with ER−/PR− tumours in postmenopausal women  and saturated fat (SFA) were associated with breast cancer risk.The study cohort consisted of 49 261 women (n=244), emigration before the follow-up (n=7), very high- or low-energy intakes <1 or 99% (n=1072) and missing information on BMI, education, use of oral contraceptives (OC), age at menarche, parity, age at first birth or breast cancer in first-degree relative (mother or sister) (n=3369). Thus, the final analysis was conducted on 44 569 women. The study was approved by the ethical committee at the University of Uppsala.Around 49 261 women returned the study questionnaires and were enrolled in the study, of whom 9% were postmenopausal. We excluded 4692 women from the initial cohort due to breast cancer before enrolment  and PR status (PR− or PR+), determined by means of an Abbott immunoassay  and SFA as continuous linear variables excluding total fat. We used the standard method for energy adjustment (vs time (The models above examine the effect of substitution of fat (or fat subtypes) for carbohydrate or protein (or fat subtype) when non-alcohol energy intake is held constant. In an additional model, we also adjusted for protein and carbohydrates by fitting the following variables: non-nutritional covariates, alcohol, protein, carbohydrate, PUFA, MUFA and total fat. It examines the effect of replacing SFA with PUFA or MUFA while keeping carbohydrate, protein and total fat (and thus energy) constant. We checked the proportional hazards assumption by plotting the Schoenfeld residuals vs time .We also tested if the associations with total fat, MUFA, PUFA and SFA differed by ER and PR status. We did not conduct any analyses with joint ER and PR status because there was a strong correlation between ER and PR status.www.r-project.org).We do not have information about menopausal status after the start of follow-up. Based on the average age at menopause in Sweden being 50 years , while having an affected first-degree relative showed an increased risk .Established non-nutritional risk factors were generally evident and, for example, parous had a lower risk than nulliparous women  leaving 30 132 women with 542 breast cancer events for the analyses after age 50 years.Analysed as a continuous variable, the HR for MUFA intake was 0.55 (95% CI: 0.28–1.09) for a 10 g/day increase, while for SFA it was 1.45 (95% CI: 0.99–2.12), and for PUFA 0.58 (95% CI: 0.32–1.05) for breast cancer after age 50 years . A lineaP=0.01). When the highest PUFA quintile was compared to the lowest, the HR was 0.54 (95% CI 0.35–0.85). When we finally combined MUFA and PUFA into unsaturated fat, the HR for breast cancer was 0.79 (95% CI: 0.62–1.01) for unsaturated fat for a 10 g/day increase and 1.21 (95% CI: 1.00–1.45) for saturated fat for a 10 g/day increase. When we adjusted for protein and carbohydrate in our additional model, lower but nonsignificant breast cancer risks were found when replacing SFA with PUFA  or MUFA  for a 10 g/day increase.Compared to women in the first quintile of MUFA intake, women in the fifth quintile had a statistically significant decreased breast cancer risk  for breast cancer after 50 years of age. There was a statistically significant trend for a decreased risk across the quintiles of MUFA intake (vs ⩾50 years) and intake of MUFA, PUFA, SFA as well as unsaturated fat (PUFA+MUFA). A statistical significant interaction with age at diagnosis (<50 vs ⩾50 years) was found for intakes of PUFA (P=0.049), SFA (P=0.048) and MUFA+PUFA (P=0.040), but not for MUFA (P=0.116). These results suggest that the effects of consumption of different types of dietary fat are mainly observed in breast cancer diagnosed after age 50 years.The point estimates in For the 974 breast cancer cases, 559 (57%) were ER+, 160 (17%) were ER− and 255 (26%) had unknown ER tumour status. With respect to progesterone receptors, 474 (49%) were PR+, 235 (24%) PR− and 265 (27%) had unknown PR tumour status. A total of 94 cases had unknown ER and PR as they were registered during 2005 when no ER/PR status details were available to us. The distribution of ER+, ER−, PR+ and PR− was similar across MUFA, PUFA, SFA and total fat quintiles.vs ⩾50) for ER+ tumours (P=0.72) or PR+ tumours (P=0.85). No other statistically significant associations between total fat, MUFA, PUFA and SFA intake and risk for ER+, PR+, ER− or PR− tumours were found in the entire cohort or before or after age 50 years, irrespective of the models used (data not shown).Overall, compared to women in the first quintile of PUFA intake, women in the fifth quintile were at a decreased risk of developing ER+ tumours  and PR+ tumours  when considering the entire cohort. Similar associations were found in the fifth quintile of PUFA intake compared to the first quintile for risk after age 50 years  for ER+ tumours and: 0.52, 95% CI: 0.28–0.98) for PR+ tumours. However, there was no significant interaction between PUFA intake and age at diagnosis .The most important limitation of our study was that dietary intake was assessed only once, involving misclassification among those who changed their dietary pattern during follow-up. Because this is likely to be nondifferential, it would attenuate the strength of any true association. However, there is a potential risk that the degree of this misclassification increases over time because women change their dietary habits. Thus, the longer the follow-up, the more attenuated the association might be and we found some evidence for this: thus among women over age 50 years, the risk associated with MUFA (as a linear continuous variable) was 0.35 (95% CI: 0.13–0.97) for follow-up 0–10 years, but 1.00 (95% CI: 0.40–2.54) for longer follow-up; there was no significant interaction between MUFA and follow-up time (n-3 PUFA if these, as reported, reduce the incidence of ER− breast tumours (Women with a high PUFA intake had a decreased risk of developing ER+ and PR+ tumours after age 50 years, while no other association with fat or its subtypes differed by ER or PR status. An increased risk for ER+/PR+ tumours with higher total fat (High fat intake may increase circulating oestrogens levels . Since PStrengths of our study include its prospective design, large size and complete follow-up. Cancer registration in Sweden is obligatory, making the assessments of cases virtually complete. Furthermore, we were able to adjust for several risk factors for breast cancer. Misclassification of fat intakes due to measurement error in the food frequency questionnaire is unavoidable, but given the study design likely to be nondifferential, and thus attenuating any true association. A recent study found a null association with total fat intake using food frequency questionnaire data but a positive association when a 7-day food diary was used (Our study provides no evidence that total fat, MUFA, PUFA or SFA is associated with overall breast cancer risk, but does not rule out the possibility that types of fat may have differential effects on breast cancer risk before and after age 50 years.


The aim of this multi-centre UK study is to examine the attributes of a cohort offered predictive genetic testing for breast/ovarian cancer predisposition. Participants are adults unaffected with cancer from families with a known BRCA1/2 mutation. This is the first large multi-centre study of this population in the UK. The study evaluates mental health, perceived risk of developing cancer, preferred risk management options, and motivation for genetic testing. Participants were assessed when coming forward for genetic counselling prior to proceeding to genetic testing. Three hundred and twelve individuals, 76% of whom are female, from nine UK centres participated in the study. There are no gender differences in rates of psychiatric morbidity. Younger women (<50 years) are more worried about developing cancer than older women. Few women provide accurate figures for the population risk of breast (37%) or ovarian (6%) cancer but most think that they are at higher risk of developing breast (88%) and ovarian (69%) cancer than the average woman. Cancer related worry is not associated with perceived risk or uptake of risk management options except breast self-examination. The findings indicate that younger women may be particularly vulnerable at the time of the offer of a predictive genetic test.British Journal of Cancer (2002) 86, 1209–1216. DOI: 10.1038/sj/bjc/6600253www.bjcancer.comCancer Research UK© 2002  Predictive genetic testing is available to some individuals unaffected with cancer where a BRCA1/2 mutation has been identified in the family  and in their early to mid 40s (Men in families with hereditary breast/ovarian cancer (HBOC) can inherit a BRCA1/2 mutation and pass it on to their daughters and sons. Women at high risk of HBOC may bear a heavy emotional burden due to their familial experiences of cancer, high bereavement rates and their own fears of developing the disease. Some women at increased risk of HBOC who have not yet been offered a genetic test experience high levels of distress  What are the levels of mental health problems and cancer related worry in a clinical cohort offered BRCA1/2 testing?; (2) Do women accurately estimate their risks of developing breast/ovarian cancer and are these related to cancer specific worry?; (3) What risk management behaviours do women engage in (or anticipate) and is this related to cancer worry and risk perception?; and (4) What is the motivation for predictive genetic testing and does this differ for men and women?Participants included 315 adults from nine UK clinical genetic centres. Those eligible were ⩾18 years, from families in which a BRCA1/2 mutation had been identified, had no previous diagnosis of cancer, and no current mental illness likely to be exacerbated by study participation.This study was conducted as part of the baseline component of a large, prospective evaluation of BRCA1/2 testing in the UK. Eligible participants have a 50% (lower if an intervening relative has died) risk of inheriting a BRCA1/2 mutation. They are usually referred by their GP after learning of the mutation in the family. Some at risk individuals were directly informed of the availability of testing if they were already under the genetics service. They were recruited to the present study by their clinical geneticist or genetic associate/nurse between 1996 and 2000 during their clinic consultation . Written informed consent was obtained. Participants were given a questionnaire to complete and return directly to the data management centre at the Institute of Cancer Research/Royal Marsden NHS Trust. The multi-centre research ethics committee (98Jan003) and all local research ethics committees approved the study.Measures were selected for validity, reliability and prior application to the study population along with a number of study-specific questions. The following pre-validated measures were included:(GHQ28: Goldberg and Hillier 1979): a brief 28-item instrument designed to assess psychiatric disorder (cases) in non-psychiatric populations previously used with medical patients. The GHQ28 includes four sub-scales: somatic symptoms, anxiety and insomnia, social dysfunction, and severe depression. A total score on the GHQ28 (binary scoring) ranges from 0–28. A cut-off score of ⩾5 is recommended by the test authors i.e. a score ⩾5 indicates psychiatric disorder. However, n=226 women). Maximum likelihood factor analysis explains 61% of the variance and a significant goodness of fit test χ2 (df=9)=25.75, P=0.003. A total score on the CWS-R ranges from 6–24. A high score indicates greater worry. No clinical cut-offs are currently available.CWS: : this fo(IES: Perceived risk of developing breast/ovarian cancer (women only) was assessed in terms of the likelihood of developing breast/ovarian cancer  and relative risk compared to the average woman . Knowledge of general population risk of breast/ovarian cancer was assessed by the question ‘Do you know what the average woman's risk is of breast/ovarian cancer throughout her lifetime?’ with response options ‘No’, ‘Yes – please specify’.Women were asked to indicate their current risk management practices and options they might consider should they be gene carriers. These included: mammography, chemo-prevention , prophylactic surgery, ultrasound, CBE, BrSE or any other screening for cancer. For BrSE women were asked about frequency. Barriers to screening were assessed using a 5-point scale from ‘Not difficult’ to ‘Very difficult’ including: physical discomfort, fear of examination, transport to screening clinic, distress caused by screening, and taking time from work/family/social obligations to attend (Study specific questions asked about reasons for wanting a genetic test (multiple choice) and whether participants considered themselves to be at increased risk of other cancers. Expectations about gene carrier status were assessed  and means of referral to the genetics clinic.2) with Yates correction. For ordered categorical variables the Mann–Whitney (MW) test for trend was used. Age was analysed as a continuous variable and, in addition, subjects were divided into three age groups . These three groups were chosen as risk management options vary for women in each group. Women under 35 years are unlikely to receive a mammogram and women over 50 years of age receive regular mammograms as part of the UK National Screening Programme.The association between categorical variables was examined by means of Fisher's exact test or the chi-squared test  and standard deviation (s.d.) or median and range . Groups were compared using analysis of variance or the Kruskal-Wallis (KW) test. Associations between psychological scores were summarised by the Pearson or Spearman correlation coefficients. The influence of a number of factors in predicting whether women practised BrSE was investigated in a multivariate logistic regression analysis. Subjects with missing data were omitted from the respective analyses.Three hundred and fifteen participants met study entry criteria; three declined (response rate 99%). Six participants were retrospectively identified as decliners of genetic testing (i.e. following their genetic counselling session) and therefore did not receive a questionnaire. Nine individuals did not return their questionnaire leaving a total of 298 (97%) completed questionnaires received .P=0.02; MW) i.e. more of the London/Sutton participants attended college or university than the Southampton participants. Participants from the six smaller centres (n=57) are younger  and have younger daughters  than those from the three larger centres.Participants at each centre were compared on demographic variables. Three centres  accounted for 81% of the participants. Two of the smaller centres (Cambridge and Birmingham) only recruited women. There are no differences between the three larger centres  except for London/Sutton patient's having a higher level of educational achievement than those from Southampton (Seventy-six per cent of the cohort are women. Eighty-two per cent of participants are married or living with a partner. Over one third have a college or university education. Most men (75%) and women (66%) are currently employed. The median age for women is 41 (21–72 years), and 48 (22–86 years) for men. In the three age groups: 53 women and 14 men are <35 years; 120 women and 24 men 35–49 years; 54 women and 33 men ⩾50 years. Most participants have children (87%). The median age of offspring is 19 (range 0–50 years). Eighty-five per cent of participants described themselves as white.P=0.7; KW), cancer worry , avoidant  or intrusive  thoughts, age  or number of children .Participants were asked whose idea it was to attend the genetics clinic: 184 (63%) reported self-referral, 42 (14%) their family's idea, 12 (4%) GP recommendation and 31 (11%) referrals came from a genetics clinic. There were no significant gender differences in referral. Type of referral was not associated with general mental health . There was no difference between the two younger groups . Forty-eight (21%) women stated that they worried about developing cancer ‘frequently’ or ‘constantly’ and 38 (17%) felt that their cancer-related worry was a ‘definite’ or ‘severe’ problem. Compared with older women, younger women worried more often  and found it more of a problem . CWS-R and GHQ28 scores were positively correlated .Younger women expressed higher levels of cancer worry  than older women   and CWS-R  scores. There was no effect of age on intrusion  or avoidance  scores. Fifty women (24% of the 205 women with an intrusion score) reported a total absence of intrusive thoughts (intrusion score=0). Thirty-six women (17% of the 207 with an avoidance score) recorded an avoidance score of 0.There were positive correlations (Pearson) between IES scales and GHQ28  had higher perceived risk of breast cancer (not ovarian) with 97% of women reporting higher than average risk compared to 81% of the other referral groups . Younger women (<50 years) have a higher perceived risk of breast  and ovarian  cancer than older women.Compared with the average woman, 197 (88%) and 153 (69%) thought they were at higher/much higher risk of developing breast and ovarian cancer respectively. Thirty (14%) and 70 (32%) women considered it not very likely that they would develop breast and ovarian cancer respectively. There was no relationship between perceived risk and GHQ28, CWS-R or IES scores. Self referred women . The corresponding figures for ovarian cancer were 14% college/university and 7% school educated (P=0.01). Age had no effect on accuracy of population ovarian cancer risk figures but younger women were more likely to give a correct figure for population breast cancer risk .Sixty and 26% of women were able to give a figure for population risk of breast and ovarian cancer respectively. Of these, only 38% gave a correct estimate for breast (defined as odds ratio of 1 in 11–15) and 8% for ovarian cancer (defined as odds ratio of 1 in 50–100). Eighteen per cent and 16% of women overestimated the population risk of breast and ovarian cancer respectively. Only 4% and 1% of women underestimated population risk of breast and ovarian cancer respectively. Higher educational status was associated with accurate figures for population breast cancer risk. Forty-nine per cent of college/university educated participants were correct compared to 33% of those who were school educated (Fifty-four (24%) women and 24 (34%) men felt that they were at increased risk of developing other cancers. Bowel , prostate (12), lung  and other gynaecological cancers (four) were the most common cancers mentioned.Table 2P=0.02) and the 35–49 groups . Two hundred and twenty-two (98%) women reported having had a cervical smear. Older women (⩾50 years) were less likely to report having had a cervical smear than younger women . Twenty-nine (13%) women reported other tests for cancer including biopsy (n=1), breast ultrasound (n=3), blood tests (n=2), CA125 (n=2), MRI (n=2), pelvic ultrasound (n=2), colonoscopy (n=4), and ‘other’ . These procedures were marginally more common in older women .These women reported having had a mammogram an average of 9 months previously (range 0 mn–10 years); a CBE an average of 7 months previously (range 0 mn–21 years). One woman reported having a bilateral mastectomy 2 years before (the other two do not report a date); ovarian ultrasound an average of 5 months previously (range 0 mn–7 years); and bilateral oophorectomy an average of 3 years previously (range 7 mn–9 years). One hundred and ninety-eight (88%) women reported performing BrSE; 22 (10%) more than once a week, 49 (22%) more than once a month, 68 (30%) monthly, 51 (22%) less frequently. Older women (⩾50 years) examined their breasts less frequently than the under 35 years . A logistic regression analysis (adjusted for level of cancer worry) indicated that no other variables predicted BrSE where age, educational status, marital status, number of children, perception of the participants' own risk, their certainty of having the gene and other risk management strategies were entered as variables.Cancer-related worry was not associated with a higher level of risk management activity  other than for women performing BrSE. Women practising BrSE had a higher level of cancer related worry than those that did not . There was no relationship between cancer worry and anticipated risk management behaviour following the test result (MW). Comparing the number of women who had had the procedure, mammography was less common in younger women .If they were found to be a BRCA1/2 carrier younger women (<35 years) were more likely to consider prophylactic mastectomy than older women . ‘Fear of the examination’ was more of a barrier to screening in the younger age group . There were no other age differences.Women were asked to rate a number of factors that might make it difficult to attend for screening  or younger women . In the two younger groups, women under 35 years were more likely to endorse ‘to make decisions about having children’ . Older women were less likely to give ‘to prepare for the future’ . There were no other age differences between women. Women were more likely than men to want the test in order to prepare for the future  and to relieve uncertainty . Childless men and women were more likely to give ‘decisions about having children’ as a reason for wanting the test than those who were parents . There was no relationship between psychiatric morbidity (GHQ) and reasons for wanting the genetic test.Reasons for wanting a genetic test are presented in Table 4P= 0.03; Fisher).When asked how certain they are about having a BRCA1/2 mutation the majority (75%) were uncertain, 22% were certain, and 3% certain that they did not. More women than men were certain  a substantial proportion of women have already had prophylactic oophorectomy. It is likely that prior to the availability of genetic testing prophylactic oophorectomy was chosen as a form of risk management in this group of high risk women in the absence of carrier status information. Actual uptake of risk management options following genetic testing will be established at follow-up. In this cohort cancer worry is not associated with consideration of prophylactic surgery. This is in contrast to other studies indicating that women with high levels of general anxiety  are moreGiven that one aim of predictive testing for BRCA1/2 is to reduce mortality by regular surveillance or surgery it is important to clarify concerns and support those having difficulty engaging in appropriate risk management behaviour. The most common barrier to screening reported by the women is taking time off from their usual obligations therefore practical steps can be taken to help women overcome such hurdles such as mobile mammography units that can be used by women under 50 years where appropriate.As with This study contributes to the psycho-oncology literature by investigating a large cohort including men and consisting mainly of individuals that have not come from well-researched families but from current clinical practice. We believe that this cohort represents a population of individuals now coming forward for BRCA1/2 testing who are new to this process. Most participants were recruited from three UK centres with established genetics services. Other centres were setting up their services during the recruitment phase of the study. Most participants were pre-menopausal women. This reflects findings from other studies (In summary, these data illustrate that while general mental health is not adversely affected by the prospect of predictive genetic testing, cancer related worry is particularly prevalent amongst pre-menopausal women many of whom self-refer to genetics services. In the future genetic testing is likely to become more rapid and more widely available. Given that many of the individuals in this cohort experience frequent and troublesome cancer worry and perceive themselves to be at high risk it is important that they continue to receive genetic counselling prior to testing in order to address concerns, receive information to make informed choices and continue to be supported both medically and psychologically.


The association with low BMI, however, tended to be weaker and nonsignificant among never smokers and never drinkers.The influence of body mass index (BMI) on oral cancer risk was evaluated in 375 incident cases and 375 age–gender-matched hospital-based controls. Low BMIs at diagnosis and 2 years before diagnosis were associated with significantly elevated odds ratios (OR for BMI ⩽22  Mortality and incidence of cancers of the oral cavity and oropharynx  have been increasing in many areas of the world in recent decades  estimated at various points in life is related to oral cancer after taking into account the influence of established aetiological factors.The present case–control study is part of an international study on oral cancer and human papillomavirus coordinated by the International Agency for Research on Cancer (IARC)  invasive cancers of the oral cavity and oropharynx, codes C 01–C 10 of the second edition of the International Classification of Diseases for Oncology , gender and hospital. One control was selected for each case in the 3-month time interval after the recruitment of the case. The overall participation among controls was 91% (375 out of 412): 91.1% in Barcelona, 91.2% in Granada and 90.5% in Seville. Exfoliated cells from oral cavity, blood samples and tissue biopsies were also collected from cases and controls.Identical questionnaires and coding manuals were used in each centre, and all interviewers received the same training and were routinely supervised. Body mass indexes were calculated from self-reported height and weight at diagnosis, 2 years before diagnosis, and in young adulthood, as weight in kilograms divided by the square of height in metres. A difference between the questionnaires used in the three centres was that participants in Barcelona were asked to report their weight at the age of 30 years, while in Granada and Sevilla they reported their weight at the age of 20 years; both are referred to herein as weight in young adulthood. Height, weight and BMI distributions were categorised into approximate tertiles (based on the whole distribution of participants) using the highest tertile as the reference category.Smoking was defined as having smoked at least one cigarette  daily for at least 1 year. Subjects were also asked about duration of the habit and amount of tobacco smoked. Drinking was defined as having drunk alcoholic beverages at least once a month, and details on amount and duration were obtained.Odds ratios (ORs) and corresponding 95% confidence intervals (CIs) were calculated for the overall number of participants using unconditional multiple logistic regression models  that incP-values were derived from two-sided statistical tests.Dose–response relationship with risk was evaluated by treating categorical ordinal variables as continuous in the logistic regression models. All vs ⩾26 kg m−2) at diagnosis , and 2 years before diagnosis  after adjusting for smoking, drinking and the intake of fruit and vegetables. Conversely, low BMI in young adulthood (⩽22 vs ⩾25 kg m−2) was not significantly related to oral cancer risk .P<0.001) and 2 years before diagnosis  among current smokers (P=0.041). A similar relationship was observed in former smokers only for BMI at diagnosis , but was not present in never smokers.The same variables are re-evaluated in P<0.001), and 2 years before diagnosis  were related to an increased risk of oral cancer among ever-drinkers, but not among never drinkers (Low BMI at diagnosis (OR: 3.64; 95% CI: 2.18–6.05; drinkers .Our results are consistent with some previous studies in showing an inverse association between leanness and increased oral cancer risk (As with most case–control studies, our study may be affected by bias and confounding. In particular, the selection of hospital controls is open to criticism. However, we chose as controls individuals who, in the event of a cancer diagnosis, would have been admitted to the same hospitals where the cases had been identified . AnotherWe observed a gradual decrease in the magnitude of oral cancer risk from current smokers to former smokers and to nonsmokers. In fact, the influence of leanness on risk lacked significance in nonsmokers, and it was statistically significant in former smokers only at diagnosis, but not 2 years before diagnosis. In contrast, among current smokers, leanness was strongly associated with an increased risk at any one point in time, suggesting that smokers, in addition to being exposed to a high level of carcinogens, may suffer from weight loss as an expression of nutritional deficiency .Similarly, it has been shown that ethanol may contribute to this effect by altering absorption and metabolism of many different nutrients . Additio


Sir,We read with much interest the paper by et al mentioned that their findings were consistent with our study showing that the risk for breast cancer among female relatives seems to be restricted to the subgroup of obligate carriers  and using the correction for genotyping as proposed in Olsen a priori probabilities (SIR=4.48) or the corrected mixed approach (SIR=5.13)  led to a point estimate of breast cancer risk among carriers lower than that calculated using either the IR=5.13) . Moreoveh cancer . When wehod used .et al, the association with breast cancer in our study appeared particularly strong in the group of mothers compared to aunts or grandmothers, even after accounting for their 50% probability of being a carrier. We estimated an SIR of 7.1 (95% CI: 1.4–21)  and the corrected mixed approach gave an increased, but not significant risk of 2.9 (95% CI: 0.04–16) (Similar to what was seen by Olsen  1.4–21)  and 2, w0.04–16) . None ofet al for a strong association with breast cancer in the group of mothers. When using an a priori probability approach, our findings were also consistent with the existence of a possibly weaker association in the group of carrier relatives other than mothers, and with the existence of a gradient in breast cancer risk with increasing probability of being a mutation carrier. Due to the small group sizes, it is not clear whether the association found in mothers was different from that found in carrier relatives other than mothers. Both retrospective and prospective international studies could help to determine whether or not mothers of AT patients have a higher risk of breast cancer than that conferred by being an AT heterozygote.Our findings are consistent with those of Olsen


A protective effect of fruits and vegetables against colorectal cancer has been supported by many epidemiologic studies. This suggests that the carotenoids frequently found in these foods play a role in the prevention of this common cancer. To examine associations between the intake of individual and total carotenoids and the risk of colorectal cancer, we analyzed prospective data from the Multiethnic Cohort Study.This analysis includes 85 898 men and 105 106 women who completed a quantitative food frequency questionnaire in 1993–1996. The participants were African Americans, Native Hawaiians, Japanese Americans, Latinos, and whites aged 45–75 years at cohort entry. After an average follow-up of 8.2 years, 1292 and 1086 incident cases of colorectal cancer were identified in men and women, respectively. Cox proportional hazard models were used to estimate relative risks of colorectal cancer.No significant associations were found between intake of individual and total carotenoids and colorectal cancer risk either in men or women, except for β-cryptoxanthin, which showed a mild protective effect in men. When the associations were investigated separately for colon and rectal cancer, lycopene intake was related to an increased risk of rectal cancer in men. A decreased risk was seen for total β-carotene in male current smokers, but the test for interaction with smoking status was not significant. No association was observed in each ethnic-sex group.Overall, our findings do not support a significant association between carotenoid intake and colorectal cancer, although some associations were seen in subgroup analyses. Several mechanisms have been proposed,4–6 which are related to the antioxidant properties of carotenoids in cancer prevention. Carotenoids also enhance gap junctional communication and increase immune response. Furthermore, some carotenoids, such as α- and β-carotene and β-cryptoxanthin, have provitamin A activity.A protective effect of fruits and vegetables against colorectal cancer has been supported by many epidemiologic studies.7–17 and recent cohort studies have reported no clear relation.18–20 In addition, randomized trials of β-carotene supplementation provided no evidence to support an effect of carotenoids on colorectal cancer prevention.21,22 The carotenoid database for U.S. foods was recently updated to include various foods23; however, few studies in North America have used this updated database to examine intakes of individual carotenoids putatively related to colorectal cancer.13,15A number of studies have investigated associations between carotenoid intake and colorectal cancer risk. Results from case-control studies were inconsistent24 We also investigated the associations stratified by race/ethnicity and by multivitamin use.To further examine associations between the intake of individual and total carotenoids and the risk of colorectal cancer, we analyzed prospective data from a large multiethnic cohort study. In addition to performing overall analyses for colorectal, colon, and rectal cancer, we examined the associations according to smoking status, because cigarette smoking can alter the circulating levels of carotenoids.25 Briefly, the cohort consists of 215 820 men and women living in Hawaii or California who are mainly from 5 racial/ethnic groups: African Americans, Native Hawaiians, Japanese Americans, Latinos, and whites. In 1993–1996, participants aged 45–75 years completed a 26-page mailed questionnaire requesting information on dietary habits, medical history, and lifestyle practices. The study was approved by the review boards of the University of Hawaii and the University of Southern California. Voluntary return of the mailed questionnaire was regarded as an expression of informed consent on the part of the participants. For this analysis, we excluded participants who were not in one of the 5 racial/ethnic groups (n = 13 992) or who provided implausible dietary data (n = 8264). We also excluded participants with a history of colorectal cancer as reported in the baseline questionnaire (n = 2154) or in the cancer registries in Hawaii or California (n = 406). Therefore, a total of 191 004 participants (85 898 men and 105 106 women) remained for the analysis.The design and implementation of the Multiethnic Cohort Study has been described in detail elsewhere.25 We performed a calibration study to assess the correlations between nutrient intakes from the QFFQ and 3 repeated 24-hour recalls for a subset of the cohort.26 Correlations for selected nutrients ranged from 0.57 to 0.74 for nutrient densities. For carotenoid densities, correlations were 0.60 for β-carotene, 0.54 for α-carotene, 0.51 for lycopene, 0.59 for β-cryptoxanthin, and 0.55 for lutein from foods. Daily intake of carotenoids, total energy, and other dietary components from the QFFQ were computed by applying a food composition table that has been developed and maintained at the Cancer Research Center of Hawaii for use in the Multiethnic Cohort Study,25 and which incorporates carotenoid values from a national database.23 Total carotenoid intake was calculated by summing the food intake of 5 major carotenoids: β-carotene, α-carotene, lycopene, β-cryptoxanthin, and lutein.Dietary intake was assessed by a quantitative food frequency questionnaire (QFFQ) that requested the frequency and amount of consumption during the past year for more than 180 food items. The QFFQ was developed from 3-day measured food records from approximately 60 men and women in each of the main ethnic groups.27 Intake of β-carotene was reported from foods and supplements separately, and as “total β-carotene,” which was the combined intake from both foods and regularly used supplements. The consumption of other carotenoids was based on foods only.In addition to the QFFQ, the baseline questionnaire asked whether multivitamins or any of 7 single vitamin and mineral supplements had been used at least weekly during the previous year. If a supplement had been taken, details about the duration and frequency for that supplement were also requested. Only regular supplement use, defined as use for >1 year, was considered for this analysis. Among the 5 carotenoids examined in this study, only β-carotene as a single supplement was covered by our questionnaire. Daily supplemental β-carotene intake was calculated based on the following choices of the approximate dosage per tablet: ≤6000 µg, 7000–15 000 µg, ≥16 000 µg, and not known. β-carotene supplement users who did not provide a dosage were assigned to the lowest category (≤6000 µg). Mean daily supplement intake was calculated by multiplying the dose by the frequency of use. Carotenoid intake from multivitamin supplements was not considered because the multivitamin brands most frequently reported in the calibration study did not contain carotenoids at that time. Correlation between supplemental β-carotene intake from the baseline questionnaire and from 3 24-hour recalls was 0.43.To identify incident colorectal cancer cases, the cohort was linked to the Surveillance, Epidemiology, and End Results (SEER) cancer registries covering Hawaii and California. The cohort was also linked to death certificate files in Hawaii and California and the National Death Index. Cancer identification and death information were complete through December 31, 2002. Colorectal cancer cases were limited to participants diagnosed with invasive adenocarcinoma of the large bowel. A total of 2378 incident cases (1292 men and 1086 women) were identified after an average follow-up of 8.2 years.We used Cox proportional hazards models with age as the time metric to calculate relative risks (RRs) and 95% confidence intervals (CIs) for colorectal cancer for men and women separately. Observation began at questionnaire completion or at age 45 for the few individuals who were younger than 45 at questionnaire completion, and ended at the earliest of these dates: date of colorectal cancer diagnosis, date of death, and date of closure .26 Carotenoid intakes were divided into quintiles determined by the sex-specific intake distribution in the cohort; the lowest quintile was used as the reference group. For β-carotene models, intakes from foods (quintiles) and supplement use (yes/no) were included in the same model so that the result for each was adjusted for the other. “Basic models” were adjusted for race/ethnicity and time since cohort entry—categorized as ≤2 years, 2–5 years, and >5 years—as strata variables and age at cohort entry as a covariate. “Multivariate models” were additionally adjusted for family history of colorectal cancer, history of intestinal polyps, pack-years of cigarette smoking, body mass index (kg/m2), hours of vigorous activity per day, use of nonsteroidal anti-inflammatory drugs, regular multivitamin use, total energy intake , alcohol intake , red meat intake (g/day), dietary fiber intake , total calcium intake , total vitamin D intake , total folate intake , and use of hormone replacement therapy (in women only). Trend tests were conducted by including a continuous variable in the model assigned the sex and racial/ethnic-specific median values within the appropriate overall quintiles.We expressed daily carotenoid intakes in terms of density  in the analyses, because we found that energy-adjusted intakes led to better correlations between QFFQ values and reference levels.P values were 2-sided.We conducted the analyses for colon and rectal cancer separately. Individuals with synchronous tumors at both sites (14 men and 5 women) were excluded from these analyses. We also investigated whether the association of carotenoid intakes with colorectal cancer risk differed by smoking status at baseline, by race/ethnic group, and by multivitamin use. Tests for interactions were based on the likelihood ratio test, comparing a model with main effects for the subgroup variables and the trend for the carotenoid and one that also includes interactive terms between carotenoids and subgroup membership. All analyses were performed with SAS software (version 9.1), and all As compared with the entire cohort, colorectal cancer cases were older, more likely to have a family history of colorectal cancer, and less physically active, both in men and women Table . In womeP for trend = 0.04). Unlike in men, the basic models in women yielded no significant association except for β-cryptoxanthin intake. In multivariate models, this association was attenuated. When we ran the multivariate models without adjusting for dietary fiber intake, which was the main confounder in the association between carotenoids and colorectal cancer, a moderate inverse association was seen for β-carotene intake among men . However, the effect of β-carotene was not as strong as that of dietary fiber because the effect of dietary fiber persisted, while that of β-carotene did not, when both of them were in the model. When we examined the associations with colon and rectal cancer risk separately in the multivariate models, lycopene intake in men was significantly associated with higher risk of rectal cancer only  .P for trend = 0.99 in multivitamin users and RR = 1.15, 95% CI = 0.89–1.50, P for trend = 0.26 in non-users; for women, RR = 1.13, 95% CI = 0.77–1.68, P for trend = 0.49 in users and RR = 1.10, 95% CI = 0.80–1.51, P for trend = 0.34 in non-users.When we ran the models for multivitamin users and non-users separately, no significant association was seen in either group: for men, RR for the highest vs. the lowest quintile of total carotenoid intake = 0.94, 95% CI = 0.65–1.36, 28,29 and epidemiologic studies have supported a beneficial effect of β-cryptoxanthin on lung cancer,30,31 but not on colorectal cancer.In this large Multiethnic Cohort Study, we found no significant associations between intake of individual and total carotenoids and colorectal cancer risk either in men or women, except for β-cryptoxanthin, which showed a mild protective effect in men. β-cryptoxanthin has been shown to decrease colon carcinogenesis in animal models,18 and the Shanghai Women’s Health Study19—did not find significant associations between carotenoid intake and colorectal cancer risk. In male smokers in the Alpha-Tocopherol, Beta-carotene Cancer Prevention (ATBC) Study, no association was seen between dietary and serum carotenoids and colorectal cancer risk.20 Furthermore, a pooled analysis of 11 cohort studies also did not support an important role for carotenoid intake in the etiology of colorectal cancer.32 Two randomized clinical trials—the Physicians’ Health Study21 and the ATBC Study22—found no beneficial or harmful effect of supplemental β-carotene on colon/colorectal cancer. A meta-analysis33 of 4 randomized trials, including the ATBC Study,34 also reported no effect of β-carotene supplementation on primary or secondary prevention of colorectal adenoma, which is the precursor of colorectal cancer.Prospective cohort studies have mainly reported the absence of an association. Two female cohort studies—the Canadian National Breast Screening Study (NBSS)10 Italy,7,8 China,12 and Japan17 found inverse associations of colorectal/colon/rectal cancer risk with intake of various carotenoids. However, studies in France,14 Canada,15 and Australia16 reported no association for intakes of different carotenoids. A study in the United Sates reported that high β-carotene intake was associated with a reduced risk for colon cancer in whites.11 Another study in the United States observed an inverse association between lutein intake and colon cancer risk in both men and women,9 and found an inverse association of lycopene intake with rectal cancer in women.13 In the present study, a higher lycopene intake was related to an increased risk of rectal cancer in women, which was unexpected.Findings from case-control studies have been inconsistent. Studies in Switzerland,24 Only a few studies have investigated associations of carotenoid intake with colorectal cancer risk according to smoking status, and the results are inconsistent. In a large cohort of Canadian women, no association was found in never, past, or current smokers.18 A pooled analysis of 11 cohort studies reported that null results were not modified by smoking status.32 A case-control study in Canada observed that a reduced risk of colon cancer was related to high intake of β-carotene among never-smokers and to high intake of lycopene among ever-smokers.15 In a case-control study conducted in the United States, lutein intake was more strongly associated with decreased risk of colon cancer in current smokers than in never smokers.9 A case-control study in Europe found the opposite association between β-carotene intake and colorectal adenoma risk in non-smokers and in past/current smokers: a protective effect of β-carotene in non-smokers, but an adverse effect in smokers.35 These 3 case-control studies and the pooled analysis examined the associations in men and women combined; no sex-specific findings were reported. We found an inverse association for total β-carotene in male current smokers, but there was no interaction between smoking status and intake of carotenoids. Therefore, this finding is likely due to chance.Oxidative stress from tobacco smoke may modify the blood concentrations of potent antioxidants such as certain carotenoids.Although only fewer than 5% of the participants in our study reported using β-carotene supplements regularly, the dosage available in supplements is usually higher than that obtainable from foods. We might have underestimated supplemental β-carotene intake at baseline because we did not record intake from multivitamins. Some multivitamin products might have contained β-carotene, although the brands most frequently used in the calibration study did not. During the follow-up period, these common products began to contain β-carotene and other carotenoids, such as lutein and lycopene. We speculate that subsequent intake of β-carotene and other carotenoids from supplements may have been considerably higher than at baseline. However, the results for multivitamin users and non-users were similar. Lack of information on colorectal cancer screening, including colonoscopy or sigmoidoscopy findings, was also a limitation, although a history of intestinal polyps was adjusted for in the multivariate models.The strengths of this study include its prospective design, the large number of participants, and detailed dietary information that was validated in the calibration study. A further strength was the ability to control for various factors that might potentially confound the associations between carotenoid intake and colorectal cancer. Most previous studies on carotenoid intake and colorectal cancer enrolled whites. However, we were able to examine the associations among various ethnic groups with a wide range of intakes from various sources, although we failed to find any difference between ethnic groups.In conclusion, our findings do not support a significant association between carotenoid intake and colorectal cancer risk.


We analyzed the claim "mammography saves lives" by calculating the life-saving absolute benefit of screening mammography in reducing breast cancer mortality in women ages 40 to 65.To calculate the absolute benefit, we first estimated the screen-free absolute death risk from breast cancer by adjusting the Surveillance, Epidemiology and End Results Program 15-year cumulative breast cancer mortality to account for the separate effects of screening mammography and improved therapy. We calculated the absolute risk reduction (reduction in absolute death risk), the number needed to screen assuming repeated screening, and the survival percentages without and with screening. We varied the relative risk reduction from 10%–30% based on the randomized trials of screening mammography. We developed additional variations of the absolute risk reduction for a screening intervention, including the average benefit of a single screen, as well as the life-saving proportion among patients with earlier cancer detection.Because the screen-free absolute death risk is approximately 1% overall but rises with age, the relative risk reduction from repeated screening mammography is about 100 times the absolute risk reduction between the starting ages of 50 and 60. Assuming a base case 20% relative risk reduction, repeated screening starting at age 50 saves about 1.8  lives over 15 years for every 1000 women screened. The number needed to screen repeatedly is 1000/1.8, or 570. The survival percentage is 99.12% without and 99.29% with screening. The average benefit of a single screening mammogram is 0.034%, or 2970 women must be screened once to save one life. Mammography saves 4.3% of screen-detectable cancer patients' lives starting at age 50. This means 23 cancers must be found starting at age 50, or 27 cancers at age 40 and 21 cancers at age 65, to save one life.The life-saving absolute benefit of screening mammography increases with age as the absolute death risk increases. The number of events needed to save one life varies depending on the prospective screening subset or reference class. Less than 5% of women with screen-detectable cancers have their lives saved. Under ideal conditions, a woman participates in screening mammography after deciding that the potential benefit , considering the limitations, outweighs the expected harms and opportunity cost (time and money) . UnrealiFurthermore, the components of informed medical decision-making regarding screening are not universally accepted -9. For eBecause popular claims such as "All women of the appropriate age should be screened" are often linked with the mantra "good quality screening mammography saves lives" , we deciAbsolute risk is defined as the probability of a particular event occurring over a specified period , such asThe second statistic, the number needed to treat to prevent one injury, is simply the reciprocal of the frequency form of the reduction in absolute risk -22. In omortality risk reduction multiplied by the absolute death risk. Therefore, the first statistic becomes the reduction in absolute death risk or RADR = RRR * absolute death risk. The second statistic is the number needed to treat to save one life, and the third statistic is the survival percentage. We also developed two additional statistics for the life-saving absolute benefit for a screening intervention. These variations include the average benefit from one screening exam (using airbags for a long trip), as well as the life-saving proportions among women who develop or have earlier detection of cancer (lives saved for those in traffic accidents). Therefore, the life-saving absolute benefit in regards to screening has five forms: the RADR, number needed to screen, and the survival percentage; as well as the average benefit and the life-saving proportions.When the event is a death, the intervention causes a life-saving absolute benefit that reduces an absolute death risk. The absolute risk reduction equals the relative absolute death risk. If we used only the 1985 higher death risk without accounting for vehicle improvements, we would overestimate the seat belt benefit. Likewise, using the 2005 lower death risk would underestimate the seat belt benefit. To get the true seat belt benefit, we need to estimate a new baseline automobile absolute death risk without seat belts but with vehicle improvements. We can derive the baseline risk by subtracting the benefit of vehicle improvements from the 1985 higher death risk.The first step in calculating the life-saving absolute benefit is to estimate the absolute death risk from breast cancer. Returning to the automobile example, assume both restraint use (seat belts) and vehicle improvements (anti-lock brakes) contributed to a reduction in the annual automobile accident absolute death risk over the last 2 decades since the promotion of seat belt use. To calculate the absolute benefit of seat belts, RADR = RRR * For breast cancer, we need to subtract the benefit of improved breast cancer treatment from the prescreening absolute death risk to get a new screen-free absolute death risk of breast cancer. Fortunately, these data are now available. Based on computer modeling, Berry et al estimated that screening mammography has been responsible for 46% (range 28%–65%) of the 30 percent breast cancer mortality reduction from 1975–2000, with the remaining 54% (range 35%–72%) due to therapy .Fletcher and Harris previously used the cumulative probability of breast cancer death to present absolute death risk ,25. We oTherefore, to estimate the current therapy but screen-free absolute death risk, we first needed to calculate a risk difference = H - L, with H = higher 1980 prescreen death risk and L = lower 2004 current death risk. We started with the SEER 1978–1980 death risk per 1000 women including both DCIS and invasive cancer, e.g. for age 40 H = 5.95  5.71–6.19). We subtracted the 2002–2004 death risk, L = 3.73 (95% CI 3.64–3.81), to obtain the risk difference or mortality improvement, H - L = 2.22. The 95% CI is 1.96–2.48 using Monte Carlo resampling. We multiplied the risk difference by the fraction attributable to therapy or 0.54. We then subtracted this therapy benefit (0.54 * 2.22 = 1.20) from the 1978–1980 death risk to obtain a new screen-free absolute death risk, e.g. 5.95 – 1.20 = 4.75 at age 40. We calculated the range (4.30–5.32) by using interval analysis.RRR * absolute death risk. The risk ratio of death from the RCT of repeated screening mammography is the ratio of cumulative follow-up breast cancer deaths in the screened group to the deaths in the control group, adjusted by the person-years in each group. The complement of the risk ratio is the RRR of repeated screening [With the new screen-free absolute death risk from breast cancer, we can calculate the life-saving absolute benefit of screening knowing the RRR since RADR = creening . Analysecreening ,29. We ccreening ,32. By mThe RRR from the analyses of the screening trials is due to women undergoing multiple or repeated screening mammograms over a median period of about 6 years. To estimate the number of mammogram examinations needed to save one life, the NNSR needs to be multiplied by the median average number of screening rounds of 4.25 (range 2–9) . Tabar eWe also estimated the fraction or proportion of women with life-saving mammography in the two following groups: those who will develop cancer, and those who can have their cancer detected by screening mammography. The numerator of the life-saving proportion (LSP) is the screen-free absolute death risk multiplied by the RRR. The denominator is the SEER 2002–2004 cumulative 15-year probability of developing cancer, multiplied by the chance of having that cancer detected (mammography sensitivity), or the cumulative cancer detection rate (CCDR). Therefore, the LSP = RADR/CCDR. We assumed a cumulative sensitivity of 80% for repeated mammography over the period of screening . The recWe can summarize our results by converting the life-saving absolute benefit of mammography in the various screening subsets that we have analyzed into the life-saving percentage of events in each subset. These prospective subsets or reference classes  include 2).To simplify our results, we also generated scatter plots of death risk and development risk versus age in Excel . We determined the best-fitting trend line by simple linear regression as defined by the highest coefficient of variation , the absolute death risk over 15 years more than doubles. Between the ages of 50 (0.883%) and 60 (1.177%), the absolute death risk is about 1%.Figure Figure The reciprocal of the RADR is the NNSR, which is 1050 (940–1160) at age 40, 570 520–620) at age 50, and 430 (400–450) at age 60. This means that if screening starts at age 50, repeated screening of 570 mostly healthy women saves one life among the many women who develop cancer over 15 years. The survival percentage shown in Table 0–620 at Table Finally, Table Using Figure We have tried to explain the purpose of mammography by answering the question: how often will screening mammography "save" a woman's life? We have shown that since the RADR = RRR * absolute death risk, the answer for women undergoing repeated screening increases with the age-related absolute death risk. For the age range 50 to 60, the screen-free absolute death risk for average women is approximately 1%, so the RRR from repeated screening is about 100 times the RADR or absolute risk reduction.The life-saving absolute benefit also depends on the prospective screening subset or reference class to which a woman will belong. For women who will develop cancer, the absolute death risk increases, so the life-saving absolute benefit of mammography increases. Women who will die of cancer have an absolute death risk of 100%, so the life-saving absolute benefit is the maximum or the RRR. For the base case age 50 and assuming a 20% RRR, women with a screen-detectable cancer have a 4.3% chance of truth for the claim that "screening mammography saved my life."We can express the life-saving absolute benefit using percentage or frequency forms. The low extreme is the average benefit from a single mammogram of 1/11050 screens (0.009% at age 40/10% RRR). The high extreme is the life-saving proportion for women with a screen-detectable cancer of 1/1370 (7.3% at age 65/30% RRR). For the base case 20% RRR, the life-saving absolute benefit of a single mammogram is 1/2970 screens, or 0.034%. For multiple or repeated screening, the RADR is 1.8/1000, so the life-saving percentage is 0.18%. Viewing these results from a different perspective of the number of events needed to save one life in each screening subset, the following values are equivalent: five women dying of breast cancer, 23 women with screen-detectable cancers, 29 women with cancers present, 570 women screened repeatedly, and 2970 women screened once. In other words, by subtracting the one life saved in each prospective screening subset, there is no life-saving absolute benefit for 22 women with screen-detectable cancers, or for 2969 women screened once.Our findings are close to other estimates of absolute benefit from screening mammography, which will vary due to the RRR and the cumulative period used. For instance, the SEER 2002–2004 10-year cumulative risk of breast cancer death from age 40 to 50 is 0.194%, versus 0.373% or 3.73/1000  over 20 years, given a 30% RRR, while the number of mammogram examinations was 1499 (95% CI 1046–2642) . This esFirst, we have quantified how the life-saving absolute benefit of mammography gradually increases with age, which is important for younger women considering the age at which to begin screening ,6. For ediagnosis risk with death risk [Our analysis of "mammography saves lives" addresses a knowledge gap and may facilitate better information exchange before women initiate screening . Many woath risk . This coath risk . SEER daath risk .Consequently, women should have improved insight and reduced anxiety with a balanced presentation of the absolute benefit of mammography in terms of the diagnosis risk as well as the death risk . Table 3One reason for the public's confusion is that screening statistics can be difficult for even highly educated physicians to understand, and this innumeracy presents a roadblock to consumer insight about the true benefit of mammography . Rather Figure Besides the opportunity cost of time and money (buying the ticket), authorities acknowledge that screening mammography harms include a 30% increase in overdiagnosis and overtreatment, delayed diagnosis, and radiation-induced cancers ,31,55. OInstead of providing professional even-handed advice, breast radiologists usually enthusiastically promote expensive new screening technologies developed by the medical imaging industry despite the drawbacks ,17,60,61some lives, but includes significant harms to many women with and without cancer. Figure with breast cancer, not from breast cancer. Education campaigns for breast cancer awareness that want to promote insight rather than uptake should stress the death risk from breast cancer along with the development risk. For example, Figure Voluntary informed participation requires that women and their physicians understand that earlier detection and treatment of breast cancer through screening mammography (and possibly through other imaging technology) certainly saves not dying from breast cancer before age 55, which improves to 99.6% with regular screening  [We have shown that between the starting ages of 50 and 60, the RADR or absolute risk reduction is only about 1% of the RRR. In order to address the problem of innumeracy, health organizations should advertise the absolute death risk and the absolute benefits along with the corresponding RRR ,68. Figuot 1.0%) .Analysts can best communicate statistical information by using natural frequencies (event counts using one reference class) or clarifying the reference class . For insA comparison with equivalent risks will help put the death risk from breast cancer in perspective. For example, the cumulative death risk from lung cancer is 0.93% or 9.3/1000 between the ages of 50 and 65 , similarFrom the RCT survival curves, analysts cannot tell if the mortality reduction associated with screening mammography is due to either a "cure" or "reduction in hazard". We have assumed that a "life saved" means screening helps cure one woman with breast cancer who would have otherwise died from the disease without screening. The end result from a cure is that a woman lives a normal life span beyond the point of expected death from breast cancer. However, all women with cancer may theoretically benefit from screening mammography through slowing the disease and therefore slightly prolonging their lives. These women would experience life extension but would still die from breast cancer. The source of the survival benefit from screening mammography may lie somewhere in between individual life saving and multiple smaller life extensions, so the distribution of women actually benefiting in some way from screening  is not known .In economic theory, the marginal cost of a defensive measure taken to reduce the risk to health and life must be less than the marginal benefit from the reduced probability of disease and death, or the action should not be undertaken . TherefoOur study is limited because the screening effect on the absolute risk of death from any cause may be more meaningful than the screening effect on breast cancer mortality from the RCT ,76. Our We have shown how the life-saving absolute benefit of screening mammography gradually increases with age as the screen-free absolute death risk increases, since RADR = RRR * absolute death risk. For 50- to 60-year-old women, the screen-free absolute death risk over 15 years is approximately 1%, so the RRR for repeated screening mammography is about 100 times the RADR or absolute risk reduction. The chance that "mammography saves lives" depends on the absolute death risk in the prospective screening subset or reference class to which a woman belongs. For a woman in the screening subset of mammography-detectable cancers, there is a less than 5% chance that a mammogram will save her life. By comparing screening mammography's life-saving absolute benefit with its expected harms and her opportunity cost, a well-informed woman along with her physician can make a reasonable decision to screen or not to screen for breast cancer.CCDR: Cumulative cancer detection rate; CI: Confidence interval; LSP: Life-saving proportion, or the reciprocal of the NND; NND: Number of cancers needed to develop or be detected; NNSO: Number needed to screen once, or the reciprocal of the average benefit; NNSR: Number needed to screen repeatedly, or the reciprocal of the RADR; RADR: Reduction in absolute death risk, or RRR times the absolute death risk; RCT: Randomized controlled trials of screening mammography; RRR: Relative risk reduction; SEER: Surveillance, Epidemiology and End Results.The authors declare that they have no competing interests.JDK conceived the study, developed the methodology, and drafted the manuscript. JEK performed the statistical analysis. Both JDK and JEK analyzed the results, and revised and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1472-6947/9/18/prepub


The 'unopposed oestrogen hypothesis' for endometrial cancer maintains that risk is increased by exposure to endogenous or exogenous oestrogen that is not opposed simultaneously by a progestagen, and that this increased risk is due to the induced mitotic activity of the endometrial cells. Investigation of the mitotic rate during the menstrual cycle shows that increases in plasma oestrogen concentration above the relatively low levels of the early follicular phase do not produce any further increase in the mitotic rate of endometrial cells. A modification of the unopposed oestrogen hypothesis which includes this upper limit in the response of endometrial cells to oestrogen is consistent with the known dose-effect relationships between endometrial cancer risk and both oestrogen replacement therapy and postmenopausal obesity; it also suggests that the mechanism by which obesity increases risk in premenopausal women involves progesterone deficiency rather than oestrogen excess, and that the protective effect of cigarette smoking may be greater in postmenopausal than in premenopausal women. Detailed analysis of the age-incidence curve for endometrial cancer in the light of this hypothesis suggests that there will be lifelong effects of even short duration use of exogenous hormones. In particular, 5 years of combination-type oral contraceptive use is likely to reduce a woman's lifetime risk of endometrial cancer by some 60%; whereas 5 years of unopposed oestrogen replacement therapy is likely to increase her subsequent lifetime risk by at least 90%; and even 5 years of 'adequately' opposed therapy is likely to increase subsequent lifetime risk by at least 50%.


IGF1) is a strong candidate gene for a breast cancer susceptibility model. We investigated a dinucleotide repeat 969 bp upstream from the transcription start site of the IGF1 gene for possible associations with plasma IGF1 levels and breast cancer risk in a multiethnic group of postmenopausal women. Furthermore, we investigated the relation between race/ethnicity, mean plasma IGF1 levels and breast cancer rates in the Hawaii/Los Angeles Multiethnic Cohort. The mean age-adjusted IGF1 level among Latino-American women, 116 ng ml−1, was statistically significantly lower than the mean age-adjusted IGF1 levels for each of the three other racial/ethnic groups, African-American, Japanese-American and Non-Latino White women  (P<0.0001). Latino-American women have the lowest breast cancer rates of any racial/ethnic group in the cohort. These results support the investigation of an expansion of the hypothesis for an important role of IGF1 in breast cancer tumorigenesis to different racial/ethnic groups and to postmenopausal women. It is unlikely that any involvement of IGF1 in breast cancer aetiology is mediated by the IGF1 dinucleotide repeat polymorphism, which was not significantly associated with circulating IGF1 levels nor breast cancer risk in this study. Research into relevant determinants of IGF1 levels in the blood must continue.The insulin-like growth factor 1 gene ( We hypothesised that the (CA)19 homozygous genotype (19/19) might indicate women with a decreased lifetime exposure to IGF1 and consequently, a decreased susceptibility to breast cancer. We examined the role of IGF1 in postmenopausal breast cancer among women from four major racial/ethnic groups  in the Hawaii and Los Angeles Multiethnic Cohort. We first examined whether the mean plasma IGF1 levels were correlated with patterns of breast cancer risk by the racial/ethnic group. We then attempted to confirm whether a microsatellite variant for IGF1 was predictive of plasma IGF1 levels in different racial/ethnic groups and to test whether the variant was a marker of breast cancer risk in our postmenopausal multiethnic population.The insulin-like growth factor 1 (IGF1) protein has been implicated in breast cancer because of its mitogenic and antiapoptotic effect on mammary epithelial cells . EpidemiParticipants included in these analyses were selected from a large multiethnic cohort (MEC) study in Hawaii and Los Angeles initiated with emphasis on diet and lifestyle characteristics in the aetiology of cancer. The cohort was established between 1993 and 1996 from driver's licence files in Hawaii and Los Angeles, including 215251 men and women, ages 45–75 years at the time of enrollment. The MEC includes primarily African Americans, Japanese Americans, Native Hawaiians, Latino-Americans and Non-Latino Whites. Baseline data were collected on all cohort members via a mailed questionnaire that contained five sections: (a) background, including medical history and family cancer history; (b) diet history; (c) medication use; (d) physical activity and (e) female reproductive history, including the use of hormones. Details of the study have been published previously  over age 55 and periods stopped; (2) 55 years or younger, periods stopped and no hysterectomy; (3) 55 years old or younger, periods stopped and bilateral oophorectomy. At the time the sample was selected , plasma was available for IGF1 testing among 123 African-American, 71 Japanese-American, 154 Latino-American and 58 Non-Latino White postmenopausal women.We tested the association between circulating plasma IGF1 levels and racial/ethnic group in 406 healthy, postmenopausal women. For a subset of these women , genotyping of the IGF1 (CA)19 variant and breast cancer risk. Women included in the case–control analysis were selected irrespective of hormone therapy status. Of the 800 women, 56 (29 cases and 27 controls) reported having a prevalent breast and/or uterine cancer on their questionnaire and were subsequently excluded from the analysis. Three women were missing weight or height values. The remaining 368 cases and 373 controls were successfully genotyped. A subset  of the controls selected for this case–control data set overlapped with controls selected for the plasma IGF1 by racial/ethnic group analysis described above.A total of 800 postmenopausal women (400 cases and 400 controls) from the same four racial/ethnic groups were selected for a nested case–control study of the microsatellite (CA)Plasma IGF1 was quantified in the Reproductive Endocrine Research Laboratory at USC, under the supervision of Dr Frank Stanczyk. IGF1 was measured by direct radioimmunoassay (RIA) using commercial kits obtained from Quest Diagnostics at the Nichols Institute . All samples were analysed in a single batch, meaning that kits with identical lot numbers were used and all assays were performed by the same individual. The intra-assay coefficients of variation for the RIA assay ranged from 9.5 to 13.8% .n variant was performed using PAGE gel electrophoresis. We performed PCR amplification of the DNA region surrounding the microsatellite repeat in question using identical primers to those used in the study by μM of each deoxynucleotide triphosphate, 2.5 μM MgCl2, 2% dimethyl sulphoxide, 1.5 U Taq polymerase  and the manufacturer's recommended buffers were combined in 25 μl reactions. The forward primer was labelled with 33P using T4 polynucleotide kinase . A ‘touchdown’ PCR cycling protocol was used which consisted of 35 cycles in total. The programme started with a 45 s denaturation at a temperature of 94°C. The first cycle continued with a 30 s annealing phase at 64°C and finished with a 30 s 72°C extension. The annealing temperature was decreased by 1°C in each of the next nine cycles, and then was maintained at 55°C for 25 cycles. Denaturation at 94°C for 45 s, and extension at 72°C for 30 s were consistent throughout the entire programme. The final extension was held for 5 min at 72°C. The radiolabelled, denatured PCR product was screened on a polyacrylamide gel by electrophoresis. Autoradiographs were exposed for 12–18 h. Two investigators scored all genotypes independently and random samples were rerun periodically to check consistency across the entire sample. The same (CA)19 homozygote control and a (CA)21 homozygote control were run on each genotyping gel. The noninformative samples were repeated in subsequent gels thereby reducing our noninformative rate to less than 1% among cases and controls.Genotyping of the microsatellite (CA)19 homozygote, thus allowing us to orient the other genotypes in relation to the (CA)19 on the PAGE gels.In order to orient the genotype information attained from the PAGE gel electrophoresis assay in terms of the number of dinucleotide repeats, we sequenced a number of homozygous samples using an ABI 3700 automatic sequencer . Two independent investigators read the sequencing output and were able to identify which sample contained the (CA)IGF1 genotype. We used a modified categorisation scheme for the IGF1 (CA)n genotype based on the paper by 19 genotype and IGF1 concentration in the blood. Means presented are least-squares means. The square root transformed plasma IGF1 levels produced the best approximate normal distribution, but the results of an analysis of the data using square-root transformations did not differ from results using data that had not been transformed. We therefore present the straightforward means derived from the nontransformed plasma IGF1 values. Odds ratios for association between the genotype and breast cancer risk were calculated using unconditional logistic regression. Odds ratios were calculated for the IGF1 genotype categorised by the number of 19 alleles (CA19) as described above. Estimates were adjusted for age and racial/ethnic group when not stratified by race. Women were categorised into four levels of age . All analyses were performed in SAS v8 .Analysis of variance (ANOVA) was used to test differences in crude and age-adjusted mean IGF1 by racial/ethnic group and −1 (crude) and 116 ng ml−1 (age-adjusted)). African-American, Japanese-American and Non-Latino White women had plasma IGF1 levels that did not differ significantly from each other, with age-adjusted means of 146, 144 and 145 ng ml−1, respectively. Further adjustment for BMI in the model did not change these results.n genotype results for a sample (N=230) of the 406 control women described above are shown in19 homozygote (19/19) is the most common genotype , but for African-American women, the 19/18 genotype is most common (20.0%) and in Japanese-Americans, 21/19 is the most common genotype (24.0%). χ2 test for differences in the distribution of genotypes across racial ethnic groups are significant for the three-level categorisation strategy presented here  and for the categorisation used by vs non-19). There is no evidence for departure from Hardy–Weinberg equilibrium in any of the racial/ethnic groups. In addition, we found no consistent statistically significant association between age and genotype among controls, cases or all women combined, providing no evidence for confounding by age.IGF1 (CA)19 genotype  IGF1 (CA)19 microsatellite and breast cancer risk. Ages of participants at blood draw ranged from 49 through 81 years. Mean age at blood draw did not differ significantly across racial/ethnic groups or between cases and controls in each racial/ethnic stratum. The pattern of body size characteristics  in these 693 women is similar to those described for the 230 control participants in the mean IGF1 plasma analysis described above. As shown in n dinucleotide repeat among controls in the case–control analysis (data not shown) is similar to the distribution shown for the controls in the plasma IGF1 analysis shown in in situ, (DCIS); N=320). Stratification by stage of disease  did not significantly alter the results from those presented here. The 19/19 genotype was not predictive of low breast cancer risk in our sample of postmenopausal African-American, Japanese-American or Latino-American women. In Non-Latino Whites, the adjusted odds ratio for breast cancer risk associated with the 19/19 homozygote as compared to the baseline non-19/non-19 genotype was 0.82, but this protective effect was not statistically significant (95% CI 0.30–2.24) (vs any-19 categories the risk effect was 1.21 (95% CI 0.83–1.75).The distribution of genotypes for the (CA)et al found the highest rates of breast cancer among African-Americans followed by Japanese-Americans, Non-Latino Whites and Latino-Americans 19 genotype as a relevant genetic marker for cumulative lifetime exposure of breast cells to IGF1. In our study population of postmenopausal women from four different racial/ethnic backgrounds, the IGF1 (CA)19 genotype was not consistently correlated with circulating IGF1 levels and thus explained neither the interindividual nor the inter-racial variation in blood IGF1 concentrations. Our result was consistent with the absence of an association in a recent study of Caucasian men . Some confidence has been gained by the fact that the distribution of allele frequencies was consistent with Hardy–Weinberg.Future work in this area should involve analysis of larger multiethnic samples, exploration for relevant determinants and markers of IGF1 levels in the blood, in order to resolve the issue of causal involvement of IGF1 in the aetiology of postmenopausal breast cancer.


P = 0.008). On multiple regression the best predictors of ‘caseness’ were high-trait anxiety, being a graduate and inaccurate risk perception. There was no difference in the level of distress shown by women with HBOC vs. those with a history of ovarian cancer only. Implications of these findings for the counselling needs of the women are discussed. The effectiveness of the clinic in improving the accuracy of risk perceptions and relieving distress is being assessed. © 2001 Cancer Research Campaign http://www.bjcancer.comOf 230 women referred to a familial ovarian cancer clinic, 196 (85%) completed a questionnaire before they attended. The data collected included pre-counselling risk perceptions and an assessment of distress. Respondents were more likely to underestimate (44%) than overestimate (19%) their risk. Those with a family history of breast and ovarian cancer (HBOC) were particularly likely to underestimate their ovarian cancer risk. The variables assessed in this study – sociodemographic, family history, distress, anxiety proneness, coping style and beliefs about health control – explained little of the observed variation in accuracy of risk perception. On the General Health Questionnaire (GHQ-30) 30% of the sample obtained scores above the cut-off (≥ 6) recommended for screening for ‘case-level’ psychological distress. Women exhibiting case-level distress were more likely to overestimate their risk (OR = 2.3). On univariate analysis low internal locus of control was associated with ‘case-level’ distress (


The risk of breast cancer associated with delivering a twin birth was examined in a population-based nested case-control study of nearly 4800 Swedish women with breast cancer and 47000 age-matched control subjects. All were aged less than 50 years and parous. After adjustment for age at first birth and parity, a 29% reduction in breast cancer risk was observed in mothers of twins relative to those who were not . These results provide evidence that women who bear twins are at reduced risk of breast cancer, one explanation for which may be their unusual levels of hormonal exposure.


The present study examined the association between abortion prior to a first livebirth and breast cancer risk among a cohort of 3,315 women who had been delivered of liveborn children between 1946 and 1965 in a group of private gynaecology practices in Connecticut and followed through 1980 for the incidence of cancer. Among women with one livebirth at the time of cohort identification, a spontaneous abortion before this livebirth was associated with a 3.5-fold increase in the risk of breast cancer. The elevation in risk was independent of some of the major risk factors of breast cancer and became more pronounced as the number of years since the abortion increased.


TCF7L2 is a transcription factor involved in Wnt/β-catenin signaling which has a variant known to be associated with risk of Type 2 diabetes and, in some studies, with risk of certain cancers, including familial breast cancer. No studies of ovarian cancer have been reported to date.TCF7L2 rs12255372 used a 5' endonuclease assay, and statistical analysis used logistic regression among participants as a whole and among a priori-defined subsets.Two clinic-based case-control studies at the Mayo Clinic were assessed including 798 breast cancer cases, 843 breast cancer controls, 391 ovarian cancer cases, and 458 ovarian cancer controls. Genotyping at No associations with risk of breast or ovarian cancer were observed . In addition, no associations were observed among sub-groups defined by age, BMI, family history, stage, grade, histology, or tumor behavior.TCF7L2 SNP, no compelling evidence for association with breast or ovarian cancer was observed.Although the biology of the Wnt/β-catenin signaling pathway and prior association between rs12255372 and numerous phenotypes warranted examination of this  TCF7L2) encodes a transcription factor involved in Wnt/β-catenin signaling pathway which encompasses an intronic single-nucleotide polymorphism  that has been associated with risk of Type 2 diabetes in linkage studies and genome-wide association studies [Transcription factor 7-like 2  were over 18 years of age and enrolled within six months of diagnosis. Invasive or borderline epithelial ovarian cancer cases (N = 391) were over 20 years of age and living in the Upper Midwest enrolled within one year of diagnosis. Controls  were frequency-matched on age and region of residence and seen for general medical examinations. Risk factor information was collected through interviews, clinical data was abstracted from medical records, and DNA was extracted from peripheral blood (Gentra). All breast cancer participants and 98% of ovarian cancer participants were white non-Hispanic. Details are provided elsewhere ,12.TCF7L2 rs12255372 was performed on 384-well plates using TaqMan™ (Applied Biosystems). For breast and ovarian cancer, respectively, call rates were high , genotypes were in Hardy-Weinberg equilibrium , white non-Hispanic control minor allele frequencies were as expected , and concordance across duplicates was 100%. Logistic regression in SAS v. 8 (SAS Institute) was used to estimate odds ratios (ORs) and 95% confidence intervals (CIs), primarily assuming an ordinal effect with simple tests for trend; heterozygous and minor allele homozygous ORs were also estimated. Ovarian cancer analyses were adjusted for age and region of residence; breast cancer analyses were additionally adjusted for oral contraceptive use, hormone therapy use, age at first birth, age at menarche, menopausal status, and smoking pack years. Stratified analyses were performed on sub-groups of menopausal status, BMI, family history, and estrogen, progesterone, and HER2 receptor status (breast cancer) and age, family history, BMI, stage, grade, histology, and tumor behavior (ovarian cancer).Genotyping at For both breast and ovarian cancer, no overall association with risk was observed . Breast cancer stratified analyses revealed no associations (p's > 0.05) among groups defined by menopausal status, BMI, family history, or tumor histology  are directly linked to carcinogenic transformation, but they are rare and mostly found in endometrioid adenocarcinomas [age scan  that alsage scan . Althougage scan , TCF7L2 ogenesis , and mutogenesis -18, indiith risk . In ovarrcinomas . Ovarianrcinomas  implicatrcinomas .Because of the importance of Wnt/β-catenin signaling in breast and ovarian cancer, TCF7L2's critical role in combining with β-catenin, clear associations of rs12255372 with non-cancer phenotypes, and suggestive evidence of association with cancer phenotypes, we hypothesized that genotypes at this SNP may be associated with risk of breast and/or ovarian cancer. However, no strong association was suggested in our analysis of almost 800 breast cancer cases and 850 controls, and almost 400 ovarian cancer cases and 450 controls.These results contrast with a smaller analysis of 592 familial breast cancer cases and 735 controls which showed a per-allele OR of 1.19 (95% CI 1.01-1.42) . Our breTCF7L2 is not likely to be associated with risk of non-familial breast cancer or ovarian cancer. Because very modest associations could not be definitively ruled out and because increased risk may exist for triple-negative breast cancer cases, much larger studies will be required for more precise estimation of rs12255372's role in cancer susceptibility. In addition, given the role of Wnt/β-catenin signaling in cancer, identified associations of SNPs in the AXIN2 and APC genes with breast cancer and the increased risk conferred by family history for these conditions even after accounting for known syndromes, additional SNPs and genes within this pathway warrant further study.Despite strong prior evidence that this SNP may play a role women's cancer susceptibility, we conclude that variation in The authors declare that they have no competing interests.ELG drafted the manuscript and oversaw the ovarian cancer study; CS initiated the collaboration and guided subset analysis; LP-O provided preliminary results and interpreted results; RAV supervised ovarian cancer analysis; ZSF performed breast cancer analysis; FSC provided key interpretive insight; KLW compiled results and performed literature review; MS performed ovarian cancer analysis; BLF oversaw ovarian cancer analysis; and FJC oversaw the breast cancer study.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2407/9/312/prepub


A case-control study was conducted over a period of 11 months in an area containing one-third of the Swedish population. One thousand and one patients participated, constituting 94% of all women newly diagnosed as having breast cancer within the area. They were compared with 1,001 age-matched, non-hospitalized controls without breast cancer, selected by paired sampling from a population register. The risk of breast cancer was slightly, but significantly, related to parity, the standardized relative risk (SRR) being 1.35 for nulliparous women as compared to ever parous. In the different parity groups a risk significantly lower than that for nulliparous women was found only for women with more than 2 children (SRR = 0.59) but the trend with parity was highly significant (P less than 0.001). Age at first birth was not found to be an important risk factor for breast cancer. SRR was lower than for nulliparous women in all groups of women with their first birth before the age of 35 years, but the difference was significant (P less than 0.05) only for those with the first birth between 20 and 24 (SSR = 0.69) and 25 and 29 (SRR = 0.69) years of age. The trend with age at first birth (P less than 0.05) disappeared after stratification for parity, suggesting that it was a confounding factor.


Hydroxymethyl glutaryl coenzyme A inhibitors (‘statins’) are carcinogenic in rodents and an increased incidence of breast cancer was reported among pravastatin users in one randomised trial. We conducted a case–control study in the General Practice Research Database to evaluate the risk of breast cancer among 50- to 79-year old women treated with statins for hyperlipidaemia. Case and control women were matched by age, general practice, duration of prescription history in the General Practice Research Database, and index date. Adjusting for history of benign breast disease, body mass index, and use of hormone replacement therapy, women currently treated with statins had an estimated relative risk for breast cancer of 1.0  compared to women without hyperlipidaemia. Untreated hyperlipidaemia was associated with an increased risk of breast cancer . The estimated relative risk among women currently receiving only non-statin lipid-lowering drugs was similar to that of women with untreated hyperlipidaemia . We found no evidence for an increasing trend in breast cancer risk with increasing duration of statin use .British Journal of Cancer (2002) 86, 1436–1439. DOI: 10.1038/sj/bjc/6600267www.bjcancer.comCancer Research UK© 2002  P=0.002) (Hydroxymethyl glutaryl coenzyme A reductase inhibitors (‘statins’) and fibrates (the other leading class of cholesterol-lowering drugs) can cause cancer in rodents . ClinicaWe used the General Practice Research Database (GPRD), an automated data source containing drug prescription and other medical information on more than 3 million residents of the UK, to study the possible relation of breast cancer to statin use and hyperlipidaemia.We conducted a matched case–control study on the risk of breast cancer in relation to use of statins, non-statin lipid-lowering agents, and untreated hyperlipidaemia among women aged 50–79 diagnosed with breast cancer in 1992–1998 in the UK General Practice Research Database (GPRD).The GPRD started in 1988 and has been used extensively for research studies; the subset of data we have used during the past decade is of consistently high quality. (n=52 924); subjects born in the same years who had a recorded diagnosis of hyperlipidaemia  but no recorded prescription for any lipid-lowering drug (n=18 189); and a random sample of 50 000 women born in the same years who had no recorded diagnosis of hyperlipidaemia and no recorded prescription for any lipid-lowering drug. We excluded women who underwent bone marrow or other organ transplantation, received cyclosporine, or had HIV infection or AIDS.The base population consisted of subjects born from 1920 to 1955 for whom at least one lipid-lowering drug prescription (see Appendix) was recorded in the GPRD , date of first recorded prescription in the GPRD (within 1 year), and general practice. Controls were required to have evidence of active follow-up in the database on the date their matched case was diagnosed (‘index date’). The same exclusions were applied to the controls as to the cases.The use of a lipid-lowering agent was ‘current’ if a subject had at least one prescription within 6 months before the index date. Past use was defined as having at least one recorded prescription more than 6 months before the index date. Subjects who were current or past users of both statins and non-statins were classed as current or past statin users only. Untreated hyperlipidaemia was defined as a recorded diagnosis of hyperlipidaemia (see ‘Base population’ for diagnostic codes) at any time before the index date in the absence of treatment.To assess the effect of duration of treatment for current users, we calculated the time interval from each subject's first prescription for a given class of drug to her last prescription plus the number of days covered by the last prescription up to the index date.We evaluated body mass index (BMI), current or past hormone replacement therapy (HRT), and history of benign breast disease as potential confounding factors. BMI was calculated from the first recorded height and weight measurement in each subject's computer record, and the measurement of weight must have preceded the index date. We considered a woman to have a history of benign breast disease if she had a diagnosis of fibrocystic disease, intraductal papilloma, or fibroadenoma recorded at any time before her index date. We defined HRT as any recorded prescriptions for systemic oestrogen (including patches but excluding creams) either alone or in combination with a progestin. We distinguished current and past users according to whether they received any HRT prescriptions within 6 months before their index date.Subjects were designated by coded GPRD record numbers and we had no access to personal identifying information. The study protocol was approved by the Scientific and Ethical Advisory Group of the GPRD.2) and had no history of benign breast disease. We report estimated odds ratios as ‘relative risks’ with 95% confidence limits. Statistical calculations were performed using SAS, version 8.01 .We used conditional logistic regression to estimate odds ratios and 95% confidence intervals. The reference group for model-adjusted analyses comprised women who had no diagnosis of hyperlipidaemia, had no recorded prescriptions for a lipid-lowering agent before their index date, did not use hormone replacement therapy, had low body mass index  was used by approximately the same proportion of cases as controls, current HRT use was more frequent among cases. Cases were more likely than controls to have a history of benign breast disease.Characteristics of the cases and controls are listed in Table 1Table 2Multivariable model-adjusted relative risks for breast cancer according to exposure group are presented in Table 3Compared to women without hyperlipidaemia, the risk was modestly increased among women with untreated hyperlipidaemia (RR 1.6 (1.1–2.5)) and among those with hyperlipidaemia currently treated with non-statin lipid-lowering drugs (RR 1.8 (0.9–3.4)). Compared with the same reference group, the RR for breast cancer among women with past statin use was similar to that of current statin users, and past non-statin users had a risk that was nearly the same as that of current non-statin users.Among current statin users, the median duration of treatment was 1.8 years and the maximum duration was 8.6 years. The data show no trend in risk with increasing duration of current statin use: the adjusted RRs of breast cancer were 1.1 (0.4–2.8) among six cases and 28 controls with current statin use of duration 3 to 5 years and 1.1 (0.4–3.0) among five cases and 26 controls with current statin use of duration longer than 5 years.2 had a slightly increased RR of 1.4 (0.9–2.1) compared to women with BMI less than 24 kg m2.Current long-term use of hormone replacement therapy (HRT) was independently associated with a cumulative increase in risk (up to RR 2.1 (1.2–3.7) for women with 24 or more prescriptions compared with non-users; see We considered that an association with hyperlipidaemia might occur spuriously if hyperlipidaemia were diagnosed incidentally during the evaluation of patients with suspicious breast lesions. Therefore, we restricted a secondary analysis to case–control sets in which the first diagnosis of hyperlipidaemia (if any) or prescription for a lipid-lowering drug (if any) was recorded 6 months or more before the index date. In this analysis of 1095 subjects (208 cases and 887 controls), the RRs among women with untreated hyperlipidaemia (1.9 (1.1–2.6)) and women who had hyperlipidaemia treated with non-statin drugs either currently (1.8 (0.9–3.9)) or in the past (2.0 (0.9–4.4)) were similar to those in the primary analysis. This additional analysis provided further evidence that there was no increased risk among women treated with statins either currently (RR 0.8 (0.5–1.4)) or in the past (RR 1.2 (0.6–2.6)).In this population-based, matched case–control study, neither current nor past use of HMG-CoA reductase inhibitors (‘statins’) showed any association with the risk of breast cancer. Longer duration of statin use was not associated with any increased risk of breast cancer.Women with hyperlipidaemia who did not use lipid-lowering drugs or who used exclusively non-statin lipid-lowering drugs had a modestly increased risk compared to women without hyperlipidaemia. This is an unexpected finding. The few epidemiologic studies that have directly evaluated the relation of hyperlipidaemia to breast cancer risk have found little evidence for any association. In a follow-up study of over 95 000 women in the western US, no increased risk of breast cancer appeared following a single elevated serum cholesterol measurement at the time of enrolment (The observed increase in risk of breast cancer related to hyperlipidaemia in our study was modest and could have arisen through unrecognized selection or information bias, unmeasured confounding, or by chance. It would therefore be inappropriate to conclude that statin treatment lowered the risk of breast cancer in our study, simply because women treated for hyperlipidaemia were at no greater risk than normolipemic women. There was no effect of the duration of current statin use on the risk of breast cancer, and the 95% confidence intervals for relative risk of breast cancer among women with untreated hyperlipidaemia and current statin users overlap considerably.As reported in several other studies (