S-transferases (GSTs). Few breast cancer studies have investigated the joint effects of multiple GSTs and a PAH biomarker.Polycyclic aromatic hydrocarbons (PAHs) may increase breast cancer risk, and the association may be modified by inherited differences in deactivation of PAH intermediates by glutathione GST gene  and the interaction with PAH–DNA adducts and cigarette smoking.We estimated the breast cancer risk associated with multiple polymorphisms in the GST polymorphisms were genotyped using polymerase chain reaction and matrix-assisted laser desorption/ionization time-of-flight assays (n = 926 of 916), PAH–DNA adduct blood levels were measured by competitive enzyme-linked immunosorbent assay (n = 873 of 941), and smoking status was assessed by in-person questionnaires (n = 943 of 973).We conducted unconditional logistic regression using data from a population-based sample of women : p = 0.43 and 0.62, respectively).Odds ratios for joint effects on breast cancer risk among women with at least three variant alleles were 1.56  for detectable PAH–DNA adducts and 0.93  for no detectable adducts; corresponding odds ratios for three or more variants were 1.18  for ever smokers and 1.44  for never smokers. Neither interaction was statistically significant (GSTT1, GSTM1, GSTP1, and GSTA1 polymorphisms to further increase breast cancer risk.We found little statistical evidence that PAHs interacted with  Polycyclic aromatic hydrocarbons (PAHs) are an environmentally ubiquitous class of compounds that are formed during incomplete combustion of organic substances. Several PAHs have been classified as probable human carcinogens . In the PAH–DNA adducts reflect not only PAH exposure but also the body’s ability to metabolize these compounds. The formation of these carcinogen–DNA adducts is recognized as a key event in the initiation of carcinogenesis . In the Inherited differences in the metabolism of PAHs, interindividual differences in enzyme expression, and induction may be key determinants in individual breast cancer susceptibility . GeneticGSTM1 null polymorphism pyrene diol epoxide, a major carcinogen present in tobacco smoke nsidered  and 3. Fco smoke . To explco smoke ; gene–enco smoke . In addico smoke  using joco smoke  and 4. Ico smoke . We alsoco smoke , but theGST variant alleles among control women only. We used unconditional logistic regression with detectable adducts as the dependent variable, and GSTs as the independent variable, controlling for age at reference and season of blood draw  for three or more variants, 0.76  for two variants, and 0.80  for one variant. When we considered the combined effects of GSTM1, -P1, -T1, and -A1 variants that are coded based on their PAH-related biologic activity (rather than for their isothiocyanate-related activity), we did not find a significant association between the number of variant GST alleles and breast cancer risk.The ORs for the association between breast cancer risk and multiple GST polymorphisms  and breast cancer risk. We observed a 56% increase in risk  among women with detectable adducts and three or more GST variants  compared with those with all common genotypes and no detectable PAH–DNA adducts. The corresponding OR for three or more variants and no detectable adducts was 0.93 . Women with one variant allele and no detectable adducts had a 44% reduced risk of breast cancer compared with women with all common alleles and no detectable adducts ; the corresponding OR for detectable adducts was not significant  but was 1.44 among never smokers . Results for the interaction between multiple GST variants and pack-years of smoking (data not shown) are similar to those shown in As shown in GST variants was 0.62  among current smokers, and 1.66  among past smokers, compared with never smokers with all four common alleles. However, the category sizes upon which these estimates are based were small (data not shown), and the p-value for the interaction on a multiplicative scale was 0.31.When we categorized ever smokers by whether they were current or past smokers, the OR for the association between breast cancer risk and three or more GSTT1, GSTM1, GSTP1, or GSTA1 polymorphism and detectable PAH–DNA adducts among the control women  or the number of variant genotypes and detectable PAH–DNA adducts . GST polymorphisms and PAH–DNA adducts among control women stratified by smoking status. Women who ever smoked and had the GSTM1 present genotype had a 35% reduced odds  of having detectable PAH–DNA adducts compared with never smokers with the same genotype. We noted similar results forever smokers with the GSTM1 null genotype compared with never smokers with the GSTM1 common genotype . Ever smokers with the GSTT1 present genotype had a 27% reduced odds of PAH–DNA adducts  compared with never smokers with the same genotype. Ever smokers with the common GSTA1 genotype had a reduced odds of PAH–DNA adducts compared with never smokers with the same genotype  or on the additive scale. We did not find statistical evidence between number of variant alleles and smoking status on the odds of developing detectable PAH–DNA adducts on the additive scale  or on the multiplicative scale . When considering the number of adducts as a continuous outcome (data not shown), we found no evidence of a main effect of the individual genotypes on the number of adducts in an age-adjusted model .We did not observe a significant association between any single GST polymorphisms and several measures of PAH exposure, including a PAH–DNA adduct biomarker. Despite the biologic plausibility of our hypothesis, we found no strong evidence for an interaction between the number of variant GST alleles with various measures of PAH exposure  and breast cancer risk, nor did we find an association among the control women between number of variant GST polymorphisms and the outcome of detectable PAH–DNA adducts. We found some evidence for a reduced risk of detectable adducts among control women who were ever smokers regardless of whether they had the GSTM1 deletion polymorphism or the present genotype .To the best of our knowledge, this is the first epidemiologic investigation of breast cancer to consider the joint effects of four genotype . We alsoGSTA1  , we did not find a significant association between the number of variant GST alleles and breast cancer risk.A few previous studies  includinGSTA1 , GSTM1 nTM1 null , and GSTTT1 null  and GSTP valine)  are mode) , although the latter may be more informative given the lack of a dose–response relationship between adducts and breast cancer risk . In our  adducts . We obsene study , while b1 status . HoweverIn an epidemiologic study, the use of a biomarker of exposure, such as PAH–DNA adducts, is considered superior because exposure assessment is not based on self-report, eliminating concerns for recall bias. Reliability in the measurement of genotype and adducts was high. Measurement error was possible, although unlikely with genotyping and analysis of adducts; however, because the laboratory was blinded to status of samples, any measurement error in measuring genotype and of PAH–DNA adduct status was independent of case–control status. Therefore, any resulting bias would have been nondifferential.Our study power was limited when we examined interactions within subgroups of women, such as those with PAH–DNA adducts. However, the LIBCSP is the largest study conducted to date with information on PAH–DNA adducts with which to evaluate these associations, and thus our study power was better than that of previously reported studies.GST polymorphisms because of their role in phase II metabolism, specifically metabolism of PAH-reactive intermediates. Because PAH–DNA adducts may be related to inherited differences in the metabolism of PAHs, this pathway was hypothesized to be important in individual breast cancer susceptibility. GSTP1 and GSTM1 have been shown to have a role in detoxification of PAH carcinogenic intermediates produced by cytochrome P450 in phase I metabolism (GSTT1 and GSTA1 have also been implicated in the metabolism of PAH intermediates (We selected the tabolism . GSTT1 amediates .GST genes (Our previous analyses have shown that breast cancer risk was modestly elevated in relation to PAH–DNA adducts , 2004b, ST genes . In this


Sulfotransferase (SULT) plays an important role in the formation of estrogen which is usually conferred as a risk factor for breast cancer. Polymorphism of the SULT1A1 may be closely associated with breast cancer. However, studies on the association between polymorphism and breast cancer have yielded inconsistent results. We performed a meta-analysis including ethnic subgroup and menopausal statue subgroup to investigate the association of SULT1A1 Arg213His polymorphism with breast cancer.PubMed, EBSCO and Web of Science databases were searched for the correlative articles up to January 2010 (10362 breast cancer patients and 14250 controls). The risk  was used to estimate the association between SULT1A1 polymorphism and breast cancer risk. All of the data from each study use either fixed-effects or random-effects.P = 0.164), but it did increase the risk of breast cancer among postmenopausal women in the dominant model . No similar effect was found among premenopausal breast cancer women . There was a significant increase in breast cancer risk among Asian women  but not Caucasian women in recessive model. There was publication bias among postmenopausal women subgroup (P = 0.002), however by using the trim and fill method, if the publication bias was the only source of the funnel plot asymmetry, it needed two more studies to be symmetrical. The value of Log OR did not change too much after the adjustment and the fail-safe number of missing studies that would bring the P-value changed was 17.We found that SULT1A1 Arg213His had no exact effect to increase the risk of breast cancer (OR = 1.07, 95% CI: 0.97-1.17, We concluded that the polymorphism of SULT1A1 Arg213His might be one of the high risk factors for breast cancer in Asian women and in postmenopausal women for all races. We should point out that the publication bias among postmenopausal women may partly account for the result, but the conclusion might not affected deeply by the publication bias. Estrogen stimulation plays an important role in human breast cancer cell growth and development. It was reported that estrogen could affect breast cancer risk through stimulating cellular proliferation and promoting tumor progression. It mighEnzymatic mechanism involves in the formation of estrogen including two main pathways. One is the sulfatase pathway which involves conversion of inactive estrone sulfate into active estrone. SulfotrThere are three families of SULTs. They are SULT1 family which is the major "phenol" SULT, sulfating a wide range of substrates including eight subfamilies, SULT2 family and SULT4 family. SULT1A1 gene locates in chromosome 16p11.2 - p12.1. Previous study reported that exon 7 of the SULT1A1 gene contained a G to A transition at codon 213 and showed that relevant polymorphism significantly reduced its enzymatic activity [For the above reasons, genetic studies of SULT polymorphisms may improve our understanding of the mechanism of SULT and enable us to screen for individuals at high risk for different cancers. However a number of studies with conflicting outcomes have been conducted on SULT polymorphism among different cancers such as lung, ovarian, prostate and bladder -10. BesiHere we performed an updated meta-analysis which was specialized in breast cancer, including 16 studies with a subgroup analysis based on ethnicity and menopausal status, using Arg/Arg vs His/His, Arg/Arg vs Arg/His, dominant model (Arg/His+His/His vs Arg/Arg) and recessive model (His/His vs Arg/Arg+Arg/His).Two investigators (Yiwei Jang and Liheng Zhou) independently obtained relevant articles through searches of PubMed, EBSCO and Web of Science databases using the following words: 'sulfotransferase or SULT', 'polymorphism' and 'breast cancer'. Studies had been case-control design and based on SULT1A1 Arg213His polymorphism either alone or in combination with other genes and the language of publication was restricted to English. All of the studies required study design, publication, breast cancer cases, controls selection and genotyping methods. We excluded articles on only breast cancer patients or on healthy persons and one case-series study. In the end, 10362 breast cancer patients and 14250 controls from 16 case-control studies were selected for this meta-analysis.The following data were collected from each included studies: first authors, year of publications, study population , sources of controls, menopausal status and the number of different genotype in all subjects.P-value of < 0.50 or I-square > 50% [P values.The risk  was used to estimate the association between SULT1A1 polymorphism and breast cancer risk, using Arg/Arg vs His/Arg, Arg/Arg vs His/His, dominant model (Arg/His+His/His vs Arg/Arg) and recessive model (His/His vs Arg/Arg+Arg/His). For each study, the between-study heterogeneity was assessed across by the chi-square based Q statistics and I-square test. Heterogeneity was considered at either a re > 50% . All of Based on the search strategy, 16 studies were selected. There are 8 studies focused on the menopausal status. All of the studies were divided into four ethnic categories: Asian, Caucasian, African and others. The study details are shown in the table P = 0.012), so we used the random-effect model to analyze the data and found that there was no relationship between Arg/His+His/His genotype and the risk of breast cancer . In the recessive model (His/His vs Arg/Arg+ Arg/His), there was no between-study heterogeneity in the odds ratios (ORs) of the studies (Heterogeneity chi-squared = 18.25 (d.f. = 12) I-squared = 34.3%, P = 0.108). Through the fixed-effect model we found that it was no relationship with breast cancer risk . We used random-effect model (Heterogeneity chi-squared = 31.11 (d.f. = 14) I-squared = 55.0%, P = 0.005) to analyze Arg/Arg vs Arg/His   to analyze Arg/Arg vs His/His   and premenopausal women (Heterogeneity chi-squared = 0.73 (d.f. = 3) I-squared = 0.0%, P = 0.866), we used both M-H method and D-L method. For all the studies included in the menses subgroup (Heterogeneity chi-squared = 20.74 (d.f. = 10) I-squared = 51.8%, P = 0.023), there was also statistical significance   but not Caucasian women  I-squared = 40.7%, P = 0.120, OR = 1.02, 95% CI: 0.92-1.13, P = 0.678) , Arg/Arg vs His/His model , subgroup for population   or lymph node status (P = 0.0027) [Prolonged exposure to high level of estrogen still has been appreciated as a risk factor for breast carcinogenesis. From previous study we knew that SULT1A1 was an important enzyme in xenobiotic metabolism because it had broad substrate specificity with a high affinity for many compounds ,32, furt 0.0027) . This me 0.0027) . One reaFrom the ethnic subgroup, we found that there was significant result among the different race. SULT1A1 R213 H increased the risk of breast cancer among Asian women but not Caucasian women in recessive model (His/His vs Arg/Arg+Arg/His) which was consistent with the previous studies. Carlsten had reported the similar phenomenon for GSTM1 polymorphism which conferred a significantly increased risk of lung cancer to East Asians but not to Caucasians. The freIn the subgroup analysis of different menopausal statue, we surprisingly found that SULT1A1 polymorphism increased the risk of breast cancer among postmenopausal women but not among premenopausal women. In the Yang's research, a possible association between SULT1A1 and breast cancer risk was also suggested for postmenopausal women . HoweverThere were some limitations of this meta-analysis. First, the unavailable genotype data from some articles was the main limitation. We did everything possible to obtain the full data on the subjects, and about 75 percent of subjects involved in various ethnic populations. Lack of original data of each study may prevent more detailed analyses such as joint effects of SNP-SNP which we hope will be demonstrated by the following studies. Next, some controls were selected from benign breast disease which have potential risks of developing breast cancer might lead to misclassification. These limitations may also explain the publication bias in postmenopausal women.In a conclusion, SULT1A1 Arg213His may be associated with breast cancer risk in Asian women and postmenopausal women among all races, although there are no exact effects to increase the risk of breast cancer in premenopausal women. Due to the publication bias we found, it encourages more studies to pay attention on the menopausal statue in further researches.The authors declare that they have no competing interests.JSL is responsible for editorial correspondence and has contributed to the conception and design of the study, the analysis and interpretation of data, the revision of the article as well as final approval of the version to be submitted. YWJ and LHZ participated in the design of the study, performed the statistical analysis, searched and selected the trials, drafted and revised the article. TTY participated in the design of the study and helped to revise the article. ZZS conceived of the study, and participated in its design and coordination. ZMS conceived of the study, and participated in its design and coordination. All authors read and approved the final version of the manuscript.


Reproductive factors are known to be aetiologically important in breast cancer, but less is known regarding their effect on breast cancer prognosis. We have investigated the prognostic effect of age at first birth and total parity using data from the Danish Breast Cancer Cooperative Group that, since 1977, has collected population-based information on tumour characteristics, treatment regimes and follow-up status on Danish women with breast cancer. Details of pregnancy history were added from the Danish Civil Registration System and the National Birth Registry. Included in the study were 10,703 women with primary breast cancer. After adjusting for age and stage of disease , the number of full-term pregnancies was found without prognostic value. However, women with primary childbirth between 20 and 29 years experienced a significantly reduced risk of death compared with women with primary childbirth below the age of 20 years . Further adjustment for oestrogen receptor status did not influence these results. The effect was not modified by age at diagnosis, tumour size or nodal status. In conclusion, low age at first childbirth, but not parity, was associated with a poor prognosis of breast cancer. We speculate whether women who develop breast cancer despite an early first full-term pregnancy might represent a selected group with a more malignant disease.


Left-handers had higher risk of breast cancer than right-handers and the effect was greater for post-menopausal breast cancer .Left-handedness may be an indicator of intrauterine exposure to oestrogens, which may increase the risk of breast cancer. Women ( Intrauterine exposure to oestrogens has been hypothesised as being a risk factor for breast cancer . Left-haBetween 1966 and 1981, the Australian town of Busselton was the site of triennial cross-sectional health surveys consisting of a general health questionnaire (including a question on handedness) and various health-related tests. In the 1981 survey, 3940 adults participated (64% of those eligible), with 14 subsequently excluded owing to a diagnosis of malignancy before the survey date. The cohort was followed to the end of 2004. Linkage to death registrations was used to identify deaths occurring within Western Australia. Information from relatives and linkage to electoral rolls was used to identify if and when participants had left Western Australia. With cancer notification mandatory in Australia, linkage to the Western Australian Cancer Registry was used to determine the incidence of specific cancers.Relative risk estimates (HRs) for left-handedness in relation to development of breast cancer were calculated from Cox proportional hazards regression models that also adjusted for age, smoking category, body mass index, number of pregnancies and menopausal status at the 1981 survey. We also fitted two further models that limited person-time at risk to pre-menopausal years  and to post-menopausal years (over age 51 years).Among the 1786 women who participated in the survey and were not excluded owing to having missing data, there were a total of 94 cases of primary breast cancer diagnosed during the follow-up period. Of 1637 right-handers, 86 (5.3%) developed breast cancer compared with 7 (7.5%) of 93 left-handers and 1 (1.8%) of 56 women who had reported being ambidextrous. Twenty-two breast cancer cases were diagnosed before the age of 51  and 72 cases were diagnosed after the age of 51 years.After adjusting for the other variables, left-handers were more likely to develop breast cancer than right-handers, but the confidence interval was wide . The risk of developing pre-menopausal breast cancer was lower for left-handed women compared with those who were right-handed or ambidextrous, but this was not statistically significant . The risk of developing post-menopausal breast cancer was significantly higher for left-handers compared to right-handers, .These results are based on very small numbers and the CIs are wide. However, they are consistent with the previous finding from a case–control study of an increase in the risk of breast cancer which is confined to post-menopausal breast cancer (


Genetic testing for mutations in BRCA1 and BRCA2 is available in Canada for women with a significant family history of breast cancer. For the majority of tested women, a BRCA1 or BRCA2 mutation is not found, and counselling regarding breast cancer risk is based on the review of the pedigree. In this prospective study, we estimate breast cancer risks in women with a family history of breast cancer and for whom the proband tested negative for a mutation in BRCA1 or BRCA2. Families with two or more breast cancers under the age of 50 years, or with three cases of breast cancer at any age, and who tested negative for a BRCA1 or BRCA2 mutation were identified. Follow-up information on cancer status was collected on all first-degree relatives of breast cancer cases. The standardised incidence ratios (SIRs) for breast cancer were calculated by dividing the observed numbers of breast cancer by the expected numbers of breast cancers, based on the rates in the provincial cancer registries. A total of 1492 women from 365 families were included in the analyses. The 1492 first-degree relatives of breast cancer cases contributed 9109 person-years of follow-up. Sixty-five women developed breast cancer, compared to 15.2 expected number (SIR=4.3). The SIR was highest for women under the age of 40 (SIR=14.9) years and decreased with increasing age. However, the absolute risk was higher for women between the age of 50 and 70 (1% per year) years than for women between 30 and 50 (0.4% per year) years of age. There was no elevated risk for ovarian, colon or any other form of cancer. Women with a significant family history of breast cancer , but who test negative for BRCA mutations have approximately a four-fold risk of breast cancer. Women in these families may be candidates for tamoxifen chemoprevention and/or intensified breast screening with an MRI. A woman's breast cancer risk is increased if she has a first-degree relative with breast cancer at a young age, or if she has more than one relative with breast cancer . The ColFor women with a significant family history of breast cancer, genetic testing for mutations in BRCA1 and BRCA2 is available throughout Canada and elsewhere. Approximately, 25% of familial breast cancer patients carry a mutation in one of the two breast cancer genes  from each family was contacted by telephone. The family history was updated through a telephone interview. Information was collected on all female first-degree relatives of all breast cancer cases within the family. Collected data included age, cancer status, prophylactic surgery, and vital status. In the event that the proband was not familiar with the family history in sufficient detail, we asked her to identify a second person in the family who was able to provide additional details. The study received ethics approval from both participating centres.5. The expected rates for Ontario and for British Columbia were obtained from the registry data recorded in Cancer Incidence in Five Continents (volume VII). The SIR was estimated from the ratio of the observed rate to the expected rate. We estimated the cumulative incidence of cancer in the cohort of first-degree relatives by applying the observed age-specific rates to a theoretical cohort of women from 25 to 75 years of age.All first-degree relatives of women affected with breast cancer in the family were identified. These women were considered as a cohort of at-risk women, and the risk of breast and other cancers were estimated in the cohort using survival analysis. Each woman was considered to be at risk of cancer from the date of the original pedigree ascertainment to the earliest of (1) the date of the follow-up telephone call; (2) the date at which cancer was diagnosed (if diagnosed with cancer), or (3) the date of death (if deceased). For each study subject, the total number of person-years at risk was calculated. Person-years at risk were then categorised according to age group, in 5-year age groupings. The annual cancer risk was estimated for the entire cohort, for each of the 5-year groupings, by dividing the number of cancers observed by the total number of person-years at risk, and multiplying by 10Among the 474 families, there was a total of 874 breast cancers  were identified in the pedigrees at baseline. We identified 1599 first-degree relatives of the 874 women with breast cancer and who did not have breast cancer at the time of the initial ascertainment. Data were missing for 423 women. For 107 women, the proband who has tested was unaffected and they were excluded. The remaining 1492 were members of a family in which an affected woman  had undergone testing for BRCA1 and BRCA2. The analysis is restricted to these 1492 women from 365 families. The mean age of these women at the time of pedigree drawing was 48.2 years (range: 17–99) and at the time of the follow-up interview was 54.3 years (range: 24–101). The mean time from genetic testing in the proband to the time of follow-up was 6.1 years (range: 1–10). The 1492 first-degree relatives contributed 9109 person-years of follow-up.Cancer Incidence in Five Continents . The standardised incidence ratios (SIRs) for breast cancer were calculated by dividing the observed numbers of breast cancer by the expected numbers of breast cancer. A total of 65 women developed breast cancer, compared to an expected number of 15.2. The actuarial risk of breast cancer in the entire cohort was 0.7% per year, or roughly 40% over the lifetime . The annvs expected number of ovarian cancers (P=0.82) or colon cancers (P=0.79). The excess cancer risk was limited to breast cancer.Subanalyses were conducted based on the number of breast cancers in relatives . The obsIn Ontario and British Columbia, women with a strong family history of breast cancer are eligible for genetic testing for BRCA1 and BRCA2. However, for most families who exhibit the characteristics of hereditary transmission of breast cancer, a BRCA1 or BRCA2 mutation is not detected. For women in these families, questions remain about their breast cancer risk and the appropriate risk management strategies. In this study, we report that women with a family history of breast cancer  are at elevated risk of breast cancer. In many centres, a negative BRCA report is reported as ‘non-informative’ . This implies that the risk of cancer is most likely higher than those with a ‘true negative’  genetic test result. However, it is unclear to what extent a woman's cancer risk is increased when faced with a ‘non-informative’ genetic test result. Our study confirms that women with a family history of breast cancer do have an increased risk of breast cancer; however, it should be noted that there was no evidence for an increase in the risk of ovarian or any other cancer. Surveillance and preventive measures should be directed towards breast cancer.P<0.001). Again, no significantly increased risk was seen for ovarian cancer . However, age-specific risks were not presented . In addition, 23% of the probands in their study were unaffected with breast cancer. This may explain why the SIR for breast cancer reported by In an earlier study, In our study, women younger than the age of 40 years had the greatest elevation in breast cancer risk, compared to age-specific general population breast cancer risks. After the age of 40 years, the breast cancer relative risks were less extreme. Recently, there has been much interest in generating and applying polygenic risk models for risk assesment in breast cancer; these are based on developing a profile of genotypes for a number of single nucleotide polymorphisms. For example, For women with a BRCA1 or BRCA2 mutation, current options include breast screening, prophylactic surgery, and chemoprevention. However, in women from high-risk families in which a BRCA1 or BRCA2 mutation is not detected, it is unclear if the screening and risk reduction options offered to these women should be the same as for women with a BRCA1 or BRCA2 mutation. MRI has been shown to be effective in detecting small breast cancers in BRCA1 and BRCA2 mutation carriers , because the proband did not have information on their clinical and/or vital status. In general, these were more distant relatives. However, if the proband selectively included relatives with breast cancer, then this selection bias might result in a spuriously elevated SIR.In conclusion, women from high-risk breast cancer families in which a BRCA1 or BRCA2 mutation can not be found need to be counselled about their increased risk for breast cancer. Although this risk is not as high as that for BRCA1 and BRCA2 mutation carriers, it is significantly higher than the general population risk and appropriate breast screening, and prevention should be recommended.


The incidence of breast cancer among 1240 women who were treated surgically for breast hypertrophy in Copenhagen, Denmark between 1943 and 1971 was determined and compared with age- and calendar period-specific rates for the Danish female population. A total of 32 cases of breast cancer had developed by the end of 1990; the expected number was 52.55, yielding a relative risk (RR) of 0.61 [95% confidence interval (CI) 0.42-0.86]. The greatest reduction in risk was observed for women who had 600 g or more of breast tissue removed . This suggests that the number of potential foci is important for cancer development in the female breast. In the group of women who were operated on before the age of 20, four cases of breast cancer developed, compared with 2.23 expected cases, to give an RR of 1.79, suggesting that the aetiology of their breast hypertrophy may be different from that for the rest of the group.


Mammographic screening before age 50 years is less effective than at older ages and the associated radiation risks are higher. We estimated how many breast cancer deaths could be caused and how many could be prevented by a decade of annual two-view mammographic screening starting at ages 20, 30 and 40 years, respectively, in the UK; for all women, and for women with first-degree relatives affected with breast cancer. We extrapolated from a radiation risk model to estimate the number of radiation-induced breast cancer deaths, and used results from randomised trials, which suggest a reduction in breast cancer mortality of 10–20% in women invited to screening before age 50 years, to estimate the number of deaths that could be prevented. The net change in breast cancer deaths was defined as the number of radiation-induced deaths minus the number of prevented deaths. For all women, assuming a reduction in mortality from screening of 20%, a decade of annual screening was estimated to induce more deaths than it prevents if started at age 20 years and at age 30 years . The corresponding estimate for screening starting at age 40 years was a net decrease of 0.46 deaths/1000 women screened and a zero net change assuming a 10% mortality reduction. Results for women with first-degree relatives with breast cancer were generally in the same direction but, because their background incidence rates are higher, the net increases or decreases were greater. In conclusion, our estimates suggest that a decade of annual two-view mammographic screening before age 40 years would result in a net increase in breast cancer deaths, and that starting at age 40 years could result in a material net decrease only if breast cancer mortality is reduced by about 20% or more in women screened. Although these calculations were based on a number of uncertain parameters, in general, the conclusions were not altered when these parameters were varied within a feasible range. The National Health Service Breast Screening Programme currently invites women in the UK aged 50–70 years for mammographic screening once every 3 years. Whether screening should be offered to women younger than age 50 years, particularly to those thought to be at a higher than average risk of the disease because of a family history of breast cancer, is a question that is frequently raised. However, mammographic screening before age 50 years is less effective than at older ages possibly because premenopausal women have denser breasts and because the tumours grow more rapidly .Exposure to mammographic X-rays confers a risk of radiation-induced breast cancer, which is greater the younger the women are when they are exposed . HoweverFor our calculations, we assumed that before age 50 years women would be screened annually with a two-view mammogram. Data collected on a sample of women in the UK NHS Breast Screening Programme during 2001–2002 and conversion factors for glandular dose give estimates of an average radiation dose to the glandular breast tissue of 4.5 mGy from a two-view mammogram .Estimates of the risk of radiation-induced breast cancer presented here were calculated using an excess relative risk model (ERR), which means that the risk of radiation-induced breast cancer is calculated relative to the estimated ‘underlying’ breast cancer incidence rate for the UK population. The risk model that was used was based on a pooled analysis of three cohort studies from The estimated risk of radiation-induced breast cancer mortality was then estimated from the risk of radiation-induced breast cancer incidence to enable a direct comparison with the number of deaths that could be prevented by screening. We estimated current age-specific breast cancer survival probabilities by taking the 10-year relative survival probability in England and Wales for women whose breast cancers were diagnosed between 1980 and 1985  to 2001 Results from randomised trials suggest that, at a population level, offering mammographic screening to women aged 50–69 years could reduce breast cancer mortality by 25%, which implies a 35% reduction in mortality among the women who attend regular mammographic screening . A revieIn these calculations, it was necessary to take account of the fact that the mortality reduction would only apply to deaths from cancers that were diagnosed during the screening period. To estimate the proportion of cancer deaths that would have been diagnosed in each screening decade, data on the age at breast cancer death were crossclassified by age at diagnosis for the deaths in England and Wales in 1998 . The breast cancer mortality rates that could be reduced by attending screening were then estimated by scaling the UK breast cancer mortality rates for 2001 by the probabilities that the cancer deaths at each age were diagnosed during the screening decade. The number of breast cancer deaths that could be prevented by attending screening was estimated by reducing these scaled rates by 10 and 20%, respectively.vice versa.The net change in breast cancer mortality was defined as the number of radiation-induced breast cancer deaths minus the number of deaths that could be prevented per 1000 women screened. A positive net change means that the radiation-induced risks are greater than the mortality benefits and Sensitivity analyses were also performed to evaluate the effects of uncertainties in the radiation risk models and mortality benefits on the results.These calculations were repeated for women with first-degree relatives with breast cancer by scaling the estimated underlying UK breast cancer incidence rates and mortality rates by estimates of risk ratios for breast cancer incidence and mortality for women with a family history of breast cancer, compared with women who had no family history of breast cancer. These risk ratios were taken from results of a reanalysis of epidemiological studies of familial breast cancer . For exaSince the radiation-induced cancers are likely to occur later in life, on average, than the deaths that are prevented by screening, we used current age-specific life expectancy for the UK  to also For all women, a net loss in years of life was estimated for a decade of screening starting at age 20 years, and screening starting at age 30 years was estimated to result in a net gain in years of life if a 20% mortality reduction was assumed . For a dWe performed a sensitivity analysis to investigate the effect on the estimated net change in breast cancer mortality of varying the assumptions and parameters within a ‘feasible range’ and compared the estimated net change in breast cancer mortality with the estimates from the original parameter values . This shVarying all assumptions simultaneously within their feasible ranges gives estimates of the possible extreme values of the net changes in breast cancer mortality . In partOur estimates suggest that for all women a decade of annual mammographic screening starting at age 20 years would cause more radiation-induced breast cancer deaths than it prevents, and starting at age 30 years it is unlikely to result in a net reduction in breast cancer mortality. However, a decade of annual screening starting at age 40 years could result in a material net decrease in breast cancer mortality if, among women screened, breast cancer mortality is reduced by about 20% or more. If the mortality reduction is 10% in women screened, then a decade of annual screening starting at age 40 years may have little or no net benefit. Results for women with first-degree relatives with breast cancer were generally in the same direction but, because their background incidence rates are higher, the net increases or decreases were greater.These calculations were necessarily based upon a number of assumptions and parameter estimates; however, the sensitivity analysis suggested that, for the most part, the conclusions were unlikely to be significantly altered by varying the parameters within a feasible range. One exception was for screening starting at age 40 years for all women, where the direction of the net effect was altered by varying the assumptions, however the magnitude of the net effects were modest eitherway.In this paper, we have focused on the comparison of the radiation-induced breast cancer deaths with the number of breast cancer deaths prevented by screening. However, the radiation-induced cancers are likely to occur later in life, on average, than the deaths that are prevented by screening. Therefore, we also compared the years of life lost and gained by each decade of mammographic screening. The conclusions from these analyses were generally similar to those from the analysis of numbers of deaths, with the exception of estimated small net gains in years of life for a decade of annual screening starting at age 30 years for all women if a 20% mortality reduction from screening was assumed, and also for all women for a decade of screening starting at age 40 years if a 10% mortality reduction was assumed .In our calculations we did not assume that women attend for regular screening after the specific decade of interest, because the question under investigation was the net effect of each decade of screening. The reason for this is that we do not think that screening could be recommended to a certain age group on the basis of guaranteed future screening attendance. Future screening attendance could reduce the magnitude of the risk of radiation-induced breast cancer mortality, if some of these cancers were detected by screening. Therefore, we investigated the effect on the estimated risk of radiation-induced breast cancer mortality of assuming 100% future screening attendance by increasing the breast cancer survival probabilities by 10 or 20% for screening before age 50 years and by 35% after age 50 years. For a 20% mortality reduction due to a decade of annual screening, assuming 100% future screening attendance reduced the net increase in breast cancer deaths from 0.86 to 0.62 per 1000 women screened starting at age 20 years, from 0.37 to 0.16 starting at age 30 years, and for starting at age 40 years this changed the net decrease from 0.46 to 0.64 per 1000 women screened. Therefore, even if we assumed 100% future screening attendance, this would be likely to alter the magnitude of the net change, but not the direction of the result.In our calculations we assumed women would be screened annually with a two-view mammogram. In the UK Age trial, two-view mammography was used for the first screen only, whereas subsequent rounds used single-view mammography (The estimates of the radiation-induced breast cancer risk were based on the linear no-threshold assumption,that is assuming the radiation risks are linear in dose down to very low doses and that there is no threshold dose below which there is no risk of cancer. A recent review of the available biological and epidemiological evidence concluded that there is direct epidemiological evidence of an excess cancer risk from fractionated radiation doses as low as 50 mGy (The estimates for women with a family history of breast cancer were based upon the assumption that the excess relative risk of radiation-induced breast cancer per unit dose for these women is the same as for all women, that is, that there is no supra or submultiplicative interaction between radiation exposure and a family history of breast cancer. However, BRCA-1 and BRCA-2 mutations appear to be associated with a reduction in efficiency of DNA repair, which suggests that there may be an interaction between these two risk factors, at least for this subgroup of women with a family history of breast cancer . To dateet al found no single excess relative or excess absolute risk model that adequately described the risk of radiation-induced breast cancer across all of the eight cohorts considered. For breast cancer risk estimation in general populations, they suggested the use of their pooled excess absolute risk model, which included four of the possible eight cohort studies. Formal statistical comparison of the fits of the excess relative risk and excess absolute risk models is not possible, but an informal comparison based on deviance values suggested that the excess relative risk model fitted the data marginally better. Furthermore, the assumption underlying the excess absolute risk model is that the risk of radiation-induced breast cancer is not related to the underlying breast cancer incidence rate in the population. This is equivalent to assuming that the relative increase in the risk of breast cancer associated with radiation is actually lower for women with a greater than average baseline risk of breast cancer, such as women with a family history, than it is for other women. As explained above, to date there is little reliable information on the risk of radiation-induced breast cancer specifically in women with a family history of the disease, but we do not think currently that such an assumption is justifiable. However, in the sensitivity analysis, we investigated the effect of using Preston et al's pooled excess absolute risk model for all women, and although the radiation risk estimates were somewhat lower than those estimated using the excess relative risk model, the conclusions were not materially altered (The risk model that was used for these calculations was an excess relative risk model based on a pooled analysis of three cohorts, including two cohorts of women who were exposed to multiple fluoroscopy examinations ( altered .et al also concluded that at least a 20% annual reduction in breast cancer mortality was necessary for the reduction in breast cancer mortality to outweigh the radiation risks if screening starts at age 40 years. Finally, Several previous studies have also estimated the radiation risks associated with mammographic screening of younger women. The estimates for women with a family history of breast cancer were for women with one or two affected first-degree relatives, but these estimates could also be applied to other groups of women that have a similarly increased underlying risk of breast cancer. For example, for a woman with one first-degree relative who was diagnosed with breast cancer before age 40 years, the conclusions would be similar to those we have presented for women with two affected first-degree relatives. Furthermore, although our estimates are based on breast cancer incidence and mortality data from the UK, the other parameters that were used in the calculations including the radiation risk models and relative risks for a family history of breast cancer were all based upon data from international studies. Therefore, it is likely that our findings would be broadly applicable to other Western populations with broadly similar breast cancer incidence and mortality rates.In conclusion, our estimates suggest that a decade of annual mammographic screening before age 40 years would result in a net increase in breast cancer deaths, and that starting at age 40 years could result in a material net decrease in breast cancer deaths if breast cancer mortality is reduced by about 20% or more in women screened. Although these calculations were based on a number of uncertain parameters, in general, the conclusions were not altered when the parameters were varied within a feasible range.


The most convincing evidence that a factor such as dietary fat is causally related to breast cancer would be obtained from a randomised controlled trial in which exposure to dietary fat intake was systematically varied. A limitation of randomised controlled trials of breast cancer prevention, however, is the large sample size required to detect plausible reductions in risk resulting from the intervention. We describe here experience over a period of 9 years with the use of one risk factor for breast cancer as a criterion for entry to a clinical trial of breast cancer prevention. The risk factor used was the presence of extensive densities in the breast tissue on mammography, which has been found by several investigators to be strongly associated with risk of breast cancer. Using this criterion for selection, 1800 subjects of mean age 46 years were enrolled between 1982 and 1986, and again between 1988 and the present. Throughout this period, the point estimate of annual invasive cancer incidence was approximately 6 per 1000 per year. The observed cancer incidence has been consistently 4-5 times the incidence expected from age-specific breast cancer incidence data for women living in Ontario. These data show that the selection of subjects for a clinical trial of breast cancer prevention using the criterion of extensive breast parenchymal densities does identify a group at substantially increased risk of breast cancer. Use of this criterion for the selection of subjects can substantially reduce the sample size required for a clinical trial of a preventive strategy.


During a population-based screening project for breast cancer, almost 15,000 women aged 50 years and over have provided a 12 h (overnight) sample of urine for research purposes. In 3,789 women the excretion of 11-desoxy-17-oxosteroids (DOOS) and creatinine was measured. Results were analysed in terms of urinary concentrations and of a ratio between DOOS and creatinine. Age had an effect on DOOS, creatinine and their ratio. Body weight and body surface area had an effect on creatinine excretion and therefore on the ratio. The following variables did not have an appreciable effect on the above-mentioned ratio: a family history of breast cancer, parity and age at first pregnancy, menopause and oestrogenic drugs, and parenchymal pattern of the breast as observed on the xeromammogram. Breast cancer was found at first screening in 106 out of 14,697 women. In 100 of these cases DOOS and creatinine were measured. Excretion values expressed as the ratio between the two, allowing for body surface area, did not differ materially from those of 100 age-matched controls. These results lead the authors to the conclusion that the determination of androgen metabolite excretion in women over 50 years of age is of no help in selecting a group at high risk of breast cancer.


This study sought to examine changes in psychological distress following cancer genetic counselling. Women attending a family cancer clinic completed questionnaires before their appointment and at 2 weeks, 6 months and 12 months after their appointment. Twenty-six women were at low risk of developing breast or ovarian cancer, 76 were at moderate risk, 46 were at high risk and 46 women had previously had breast or ovarian cancer. All groups were compared with regard to measures of anxiety, depression, general psychological distress, worry about developing breast and ovarian cancer, and perceived risk of developing breast/ovarian cancer and perceived likelihood of carrying a genetic mutation. General psychological distress did not change over the course of the study and the groups did not differ on these measures. Worry about developing breast cancer and perceptions of the likelihood of carrying a genetic mutation significantly reduced following genetic counselling. On the whole women who had already had breast/ovarian cancer showed more concerns about ovarian cancer and raised perceptions of risk in comparison with the other groups, indicating the need for sensitive counselling of such women.British Journal of Cancer (2002) 86, 43–50. DOI: 10.1038/sj/bjc/6600030www.bjcancer.comThe Cancer Research Campaign© 2002  The population risk of developing breast cancer in the UK is one in 12 and the risk of ovarian cancer is one in 100. A small proportion (approximately 5–10%) of women who develop breast and ovarian cancers have an inherited genetic susceptibility to these cancers . To dateWidespread publicity about the possible genetic basis of some breast and ovarian cancers has lead to an increase in concern amongst women with a family history of these cancers. In many cases this concern may be unfounded, as the vast majority of these cancers are not due to an inherited genetic predisposition. However, increasing numbers of women are attending clinics in hospitals in the UK for genetic counselling about their family history of breast or ovarian cancer where most will want information about their future risk of developing cancer  who make up a significant proportion of those being counselled, in addition the study classifies unaffected women in terms of their actual risk of developing breast cancer. Evaluation of the psychological effects of genetic counselling among affected women has been relatively neglected in the literature on psychological distress. The assumption seems to be that these women will be less distressed and worried in the face of being at increased risk as they have already had cancer and therefore received the ‘worst possible news’. The current study explores whether these women are in fact less distressed and worried than unaffected women. In recent years the counselling offered to affected women has increased considerably due to a changing emphasis from research to service and is now broadly similar to that offered to unaffected women. However, it does differ. Women who have already had cancer are counselled that, if they have a genetic predisposition to cancer, their chance of developing a second primary cancer would be increased. However, specific risk figures are not usually given . Unaffected women, on the other hand, are given an estimate of their future risk of developing cancer, with reference to epidemiological data ; to look at changes in distress and worry over time and differences among groups for any changes.n=577) were sent a questionnaire prior to this appointment and 412 women returned a completed questionnaire. However, 121 of the women who were sent an appointment failed to attend for this (a non-attendance rate of 21% which is in keeping with other clinics). The response rate to the questionnaire among attenders at the clinic was 86% (393 out of 456). At their consultation, all women were asked to participate in the research by completing questionnaires. Women were asked to participate regardless of whether or not they had completed the first questionnaire. In addition, some women attended the clinic unexpectedly with a relative, and so had not been sent a questionnaire previously as they had not been sent an appointment (n=69). As these women were also at risk, or had been affected with cancer, effort was made to include them from this stage onwards. Of the 525 women who attended the clinic during the study period 96% (505 out of 525) agreed to take part in the research and 83% of these returned a completed questionnaire in the 2 weeks following their appointment (421 out of 505). Six months later 362 women were recontacted and sent a further questionnaire. Eight-five per cent (308 out of 362) were returned. Six months after this (12 months post consultation) the final questionnaire for the study was sent to 270 women and 91% (246 out of 270) were completed. For the purposes of this study only women who had completed all four questionnaires  are included in the following analyses (n=203).Patients were recruited into the study at the Department of Clinical Genetics at Guy's Hospital between May 1997 and May 1999. Criteria for referral are shown in Table 1The 203 women had a mean age of 42.3 years . Most (73%) had a partner at the time of the study and the majority (73%) had at least one child. Ninety-seven per cent of the women classified themselves as white. Most (70%) of the women currently worked outside the home.Women with a family history of breast/ovarian cancer who attend the family cancer clinic are either seen by a doctor  or a genetic counsellor (graduate nurses or science graduates with training in genetic counselling) for non-directive counselling. The sessions take between 45 min and 1½ h. Prior to attending the clinic women will have completed a family history sheet detailing the number of cases of cancer in their family, the type of cancer, the relationship of the person to the women, the age at diagnosis and death (if applicable). During the consultation a more detailed family history is taken. Using this information the woman's personal risk of developing breast cancer is calculated. At the time of this research the method used for risk assessment was the CASH model (n=2) or they had a family history of ovarian cancer only and their risk of developing breast cancer was therefore difficult to estimate (n=4). Files were missing for three women and therefore their risk could not be confirmed. The four groups of women: affected, low risk, moderate risk and high risk will be compared in the following analyses. (Forty-eight (24%) women underwent genetic testing in the course of the study: 39 affected women had blood taken to search for a BRCA1/2 mutation and nine unaffected high risk women underwent predictive testing. All had received a test result at 6 months post-consultation, although for the affected women this was a result for screening only part of the BRCA1 gene, and 37 received a result at this time stating that nothing had been found. A mutation was found in eight women and three women received a negative result on a predictive test.) With this sample size (n=203), power calculations showed good power to detect differences of any clinically meaningful magnitude in the variables examined in the study. For example, a sample size of 203 gives 90% power, at a 5% significance level, to detect a difference of 2.3 in breast cancer worry between two of the four groups. The power for comparisons between times will be even greater because of the within-subjects design.Forty-six (24%) of the women had previously had breast and/or ovarian cancer (referred to in this study as ‘affected’ women): 41 had had breast cancer, one woman had had ovarian cancer and four women had had both. The mean time since diagnosis of breast cancer was 5 years (range 4 months to 32 years) and for ovarian cancer the mean time was 4 years (range 6 months to 13 years). One hundred and fifty-seven women were unaffected and had not previously had cancer. For the purposes of this study these women were classified into three risk groups (based on the current UK guidelines) using the risk estimate that was written in their hospital notes. Women had been told this figure and what it meant during their consultation. Twenty-six women (13.4%) were classified as at low risk (lifetime risk of developing breast cancer less than 1 in 6), 76 (39.2%) were classified as at moderate risk (lifetime risk of developing breast cancer between 1 in 6 and 1 in 4 inclusive) and 46 (23.7%) were classified as at high risk (lifetime risk of 1 in 3 or higher). Data were missing for nine women who are not, therefore, included in the following analyses of variance. Six of these women had not been given a breast cancer risk estimate at the clinic as their family history data was incomplete . The scales each include six items scored from 1 to 4 with labelled response categories, giving a possible range of scores of 6–24. The scale includes items to assess how worry about developing cancer has affected mood and activities in the last month, and the frequency and intensity of worry. Women who had already had cancer were asked to specify their worry about developing cancer again. Cronbach alpha reliabilities for the scale at the four time points were greater than 0.82.A scale designed to assess specific worry about developing cancer was used  to 2 (much more likely)). Some research =2.93, P<0.01). They were also less likely to have a partner  or to have any children . These demographic differences are likely to be connected to one another. In addition, a higher proportion of affected women had incomplete data .Independent Prior to genetic counselling the mean score for women's levels of anxiety as measured by the HADS was 6.7 (s.d. 4.0) and for depression it was 3.0 (s.d. 2.9). Taking scores above 8 as a cut off for possible anxiety or depression disorder, 41% of the sample scored above this on the anxiety subscale and 11% did so on the depression subscale. These mean HADS scores are comparable to those found in other studies of women undergoing genetic counselling for breast/ovarian cancer risk e.g.  and Watst-test showed that, as predicted, there was significantly less worry about ovarian cancer than breast cancer pre-consultation (t=15.1(188), P<0.0001). This difference persisted at 2 weeks (t=12.8(188), P<0.0001), 6 months (t=12.1(188), P<0.0001) and 12 months post consultation (t=10.0(188), P<0.0001).The pre-consultation mean score for worry about breast cancer was 12.4 (s.d. 3.4) and for worry about ovarian cancer it was 8.3 (s.d. 2.9), levels comparable to those found by The mean score for perceptions of risk of developing breast cancer was 1.2 (s.d. 0.75) and for likelihood of carrying a mutation it was 0.8 (s.d. 0.85). Examining these variables in terms of proportions of women endorsing the given options, 47% of women felt that their risk was slightly higher than other women and a further 36% felt it was much higher. For perceived likelihood of carrying a mutation, 43% of the women felt this was fairly likely and 20% extremely likely.Repeated measures analyses of variance (ANOVAs) were carried out to examine changes over time in anxiety, depression, general distress levels, specific worry about developing cancer and perceptions of risk and also to examine whether the groups differed on these measures. Group means (s.d.) at each of the four time points are shown in Table 2No significant changes were found in levels of general anxiety, depression, or psychological distress over the 12 month study period. In addition, the groups did not differ on these measures and there were no significant interaction effects .t(186)=7.18, P<0.0001) with a further smaller reduction from time 2 to the 6 month follow up (t(186)=2.59, P<0.01). This was sustained until 12 months follow up with no significant further decrease (t(186)=0.82, ns). There was no group effect or interaction effect on this measure.For the sample as a whole, there was a significant reduction in worry about breast cancer over the course of the study . A pairepost hoc test  showed that affected women were significantly more worried than women at moderate and high risk about developing ovarian cancer. Low risk women did not differ from the other groups =2.95, P<0.01), with no significant reduction occurring between times 2 and 3 (t(175)=1.79, ns) or 3 and 4 (t(175)=1.01, ns). In addition, groups differed on this measure. Post hoc tests revealed that women at low or moderate risk felt it was less likely that they carried a mutation than did women at high risk or affected women . Post hoc paired t-tests were then carried out for each group separately for changes in perceived likelihood in order to clarify the pattern of changes seen in t-test results indicate that perceptions are fairly stable among the low and moderate risk women over the course of the study, whereas changes occur among high risk and affected women. Perceptions of the likelihood of carrying a gene mutation reduces for high risk women following the counselling session (t(36)=2.95, P<0.01), then significantly increases at 6 months post consultation (t(36)=−2.71, P<0.01), and significantly decreases again at 12 months post consultation (t(36)=1.87, P<0.05). For affected women, there is no immediate reduction in perceived likelihood (t(40)=0.00, ns), then a significant reduction at 6 months post consultation (t(40)=3.42, P<0.01), followed by a significant increase in perceptions at 12 months post consultation (t(40)=−2.48, P<0.05). For low risk and moderate risk women, other than a significant reduction following counselling for low risk women (t(23)=2.81, P<0.01), perceptions remain stable. For low risk women there is no change at 6 months (t(23)=0.00, ns) or 12 months (t(23)=1.81, ns). For moderate risk women there is no change immediately after counselling (t(73)=1.00, ns), at 6 months (t(73)=1.54, ns), or 12 months (t(73)=1.59, ns). Therefore, although there is enough overall change for the ANOVA to be statistically significant for these two groups (low and moderate risk women), when the pairwise t-tests are carried out it can be seen that the individual change between time points is small or, indeed, insignificant.Over the course of the study significant changes occurred in perceptions of the likelihood of carrying a mutation . Paired ed women . Howevered women . IndividThis prospective study enabled changes in psychological distress, worry and perceptions of risk to be assessed over the course of 1 year among women at differing levels of objective risk. Prior to attending for genetic counselling, women did not have raised levels of general psychological distress, anxiety or depression in comparison to other women attending for counselling or breast screening. However, on the whole they were worried about developing breast cancer, perceived themselves to be at greater risk of breast cancer than other women their age and felt that it was likely that they carried a mutation that would increase their risk of developing cancer. Women who had already had, and been treated for, cancer were found to be particularly worried about developing ovarian cancer and perceived the likelihood of them carrying a genetic mutation to be high.There was no evidence found in this study that genetic counselling raises worry. Levels of worry about developing breast cancer in fact reduced following genetic counselling, regardless of what risk women had been given. The greatest reduction in worry occurred immediately after counselling, suggesting that the counselling had a positive influence rather than it simply being the passage of time that lead to a reduction in worry. The study allowed for a comparison between levels of worry about developing breast and ovarian cancer. Stark differences were found, with worry being much greater about developing breast cancer. There were no changes in the level of worry about developing ovarian cancer over the course of the study which may be because level of worry was fairly low at the outset , perhaps due to the rarer nature of ovarian cancer and it receiving less media coverage. It is worth noting, however, that if a mutation is found in a woman, although ovarian cancer is less likely to develop than breast cancer, in relation to population risk it is markedly increased and perhaps worry is therefore unrealistically low.et al, 2001). This study indicates that whilst the result is in fact inconclusive it is actually interpreted by most women in a positive light as indicating that no mutation is present. At 12 months post consultation there is a significant increase in perceived likelihood again, perhaps as the time since the inconclusive result has passed and anxieties begin to increase towards pre interim result levels. For high risk women perceptions of the likelihood of carrying a genetic mutation actually increase at 6 months post consultation, which may be due to the lack of genetic testing available to most of these women at the present time and therefore their concerns over their risk are not alleviated. Low and moderate risk women show little change in perceptions, which started low and remained so following counselling. The findings concerning this variable would need to be replicated in view of the fact that only one item was used to assess perceived likelihood and therefore the measure has unestablished reliability. The pattern of change in the different risk groups is, however, interesting.Prior to their consultation, taking into account the rarity of BRCA1 and BRCA2 mutations, the sample of women as a whole grossly overestimated their likely chances of being a mutation carrier. However, following the genetic counselling session perceptions of the likelihood reduced most markedly. There was a significant interaction between time and group for this variable and therefore changes over time cannot be interpreted separately from group. For affected women there was a significant decrease in perceived likelihood at 6 months. This is likely to be due to 37 of these women receiving an ‘inconclusive’ result stating that two thirds of the BRCA1 gene had been searched and no mutation had been found and that the remaining third and all of BRCA2 would be searched. A study examining the interpretation of such results has been prepared for publication .One limitation of the current study is that it was based in only one clinic therefore the sample is fairly small, which may affect the generalisability of the results. The findings therefore need to be replicated in other clinics with different counsellors. In addition, it was not possible to follow up every woman who attended the clinic and began to participate in the research for the full 12 months. The finding that those with incomplete data were younger is in line with that found by Past psychiatric history or use of psychotropic medication was not assessed. There is, however, no reason to suppose that these factors would systematically vary by group.In conclusion, then, specific worry about developing breast cancer was reduced for all groups following genetic counselling, although perceptions of risk did not change. Counselling had no impact on general levels of distress, which were not, however, appreciably raised before counselling. On the whole, women who have already had cancer showed raised levels of concern about developing ovarian cancer and felt more at risk of developing breast cancer and of having a mutation, in comparison with other groups. This indicates the need for sensitive counselling of such women.


To evaluate the impact of timing, mutation carriers who had received the DA before the test result (n=47) were compared to mutation carriers who received the DA after the test result (n=42). Data were collected on well-being, treatment choice, decision and information related outcomes. The impact of the DA was measured 4 weeks after blood sampling. The impact of timing was measured 2 weeks after a positive test result. The DA had no impact on well-being. Regarding decision related outcomes, the DA group more frequently considered prophylactic surgery (P=0.02) corroborated with higher valuations (P=0.04). No differences were found for the other decision related outcomes. Regarding information related outcomes, the DA group felt better informed (P=0.00), was more satisfied with the information (P=0.00), and showed more accurate risk perceptions. Timing of the DA had no effect on any of the outcomes. No interactions were found between the DA and history of cancer. In conclusion, women being tested for a BRCA1/2 mutation benefit from the DA on information related outcomes. Because timing had no effect, the DA is considered useful either before or after the test result.The aim of the study was to evaluate the impact of a decision aid (DA) and its timing in women being tested for a BRCA1/2 mutation. Women with and without a previous history of cancer were included after blood sampling for genetic testing. The DA consisted of a brochure and video providing information on screening and prophylactic surgery. To evaluate the impact of the DA, women were randomised to the DA group ( The discovery of the BRCA1 and BRCA2 genes has induced widespread interest in genetic testing for inherited susceptibility of breast and ovarian cancer (An important reason to ask for genetic testing is to obtain certainty about the need for screening and/or prophylactic surgery (Decision aids (DAs) are interventions designed to help people make specific and deliberative choices among options by providing information on the options and outcomes relevant to a person's health status , Groningen , and Maastricht . These clinics cover the population of the eastern part of the Netherlands. The closing date for inclusion was November 2001. Women were excluded if they had a cognitive disorder that precluded informed consent, had insufficient knowledge of the Dutch language, were diagnosed with distant metastases, had undergone both bilateral mastectomy and oophorectomy, or had been treated with chemotherapy, radiotherapy, or surgery for breast or ovarian cancer less than 1 month before blood sampling. The study was approved by the local research ethics committees.The DA consisted of a brochure and video. Unlike usual information material, our information concentrated on contrasting treatment options. The 14-page brochure presented detailed information on treatment options available in 1998 in the Netherlands, and on the physical, emotional, and social consequences in qualitative terms, but whenever possible in quantitative terms  to avoid contamination. Randomisation was computer generated in blocks of 10, and stratified by personal medical history of breast/ovarian cancer. Randomisation was performed after obtaining informed consent and the baseline assessment. Neither subjects nor members of the study staff were blinded to intervention assignment.before the test result (DA_b), and women from the control group subsequently received the DA after the positive test result (DA_a). At 2 weeks after a positive test result (T3), the impact of timing of the DA was assessed by comparing the DA_b group with the DA_a group  children, and being religiously affiliated) and medical background .anxiety     to 10 .We collected data on n Scale) , and canTreatment choice was asked for both breasts and ovaries. The treatment choice related to breast cancer risk was between ‘prophylactic mastectomy’, ‘breast cancer screening’, and ‘undecided’. The treatment choice related to ovarian cancer risk was between ‘prophylactic oophorectomy’, ‘ovarian cancer screening’, and ‘undecided’. Valuations for the treatment options were asked on a 10-point scale ranging from 1 (very bad) to 10 (excellent).The decision related outcomes were asked separately for the breasts and ovaries. An overall score was created by adding the scores for the decision related to breast and ovarian cancer risk and dividing this by the number of items included.Strength of treatment preference was asked on a 4-point Likert scale ranging from 1 (weak preference) to 4 (very strong preference). Those who filled out ‘undecided’ as treatment choice were assigned a value of 0 (no preference).Decision uncertainty was measured with three items related to the uncertainty subscale of the Decisional Conflict Scale by Preference for decision-making was measured with the two decision-making items from the Problem-Solving Decision-Making Scale (PSDM) by Subjective knowledge was measured by asking women to rate their knowledge for prophylactic mastectomy, breast cancer screening, breast self-examination, prophylactic oophorectomy, and ovarian cancer screening, on a 10-point scale ranging from 1 (very bad) to 10 (excellent). An overall score for subjective knowledge was created by adding the scores on the five items and dividing this by five.The amount of received information was measured on a 7-point scale ranging from 1 (I received way too little information), 4 (I received exactly enough information), to 7 (I received way too much information), for the decision related to breast and ovarian cancer risk. An overall score was created by adding the scores and dividing this by two.Satisfaction with quality of information was measured with a 13-item questionnaire on a 6-point scale ranging from 1 (not satisfied), 4 (rather satisfied), to 6 (very much satisfied). Women were offered a series of items regarding cancer risks, efficacy of treatment options, and physical, emotional, and social consequences. An overall score was created by adding the scores on the 13 items and dividing this by 13.Risk perception was asked for the following eight items : breast (8–14%) and ovarian cancer risk (1–3%) in the general female population, breast (60–85%) and ovarian cancer risk (15–60%) in BRCA1/2 mutation carriers, the possibility of cure when breast (65–80%) and ovarian cancer (35–50%) is detected during screening in BRCA1/2 mutation carriers, and residual risk after prophylactic surgery of breasts (3–12%) and ovaries (3–12%) in mutation carriers. Women were asked to give a risk estimate in a range from 0 to 100%. For each item, a new variable was created classifying the risk estimate as underestimate, accurate, or overestimate. In the analyses, risk accuracy was dichotomised in accurate and inaccurate.We did not assess the amount of received information, satisfaction with quality of information, and risk perception at baseline to avoid information-seeking behaviour in the control group, thus making the control group less representative.To detect a difference of at least 10% in the decision uncertainty score between the DA and control groups, with a 5% two-sided significance level and a power of 80%, we needed a sample size of 180 women in each group.2 tests for categorical variables, and t-tests for continuous variables. To evaluate the impact of the DA, we compared the DA group with the control group at T2 (see P<0.10), were also included as covariate. Effect sizes (d) were calculated as the adjusted mean of the DA group minus the adjusted mean of the control group, and as the adjusted mean of the DA_b minus the adjusted mean of the DA_a group, divided by the standard deviation of the difference score. For two categorical variables, namely ‘treatment choice’ and ‘accuracy of risk perception’, comparisons between groups were made using χ2 tests. We used a P level of 0.05 to indicate statistical significance. The number of subjects providing data for the various analyses varied due to missing data.We analysed data from women who completed the questionnaire at baseline and at the time point of interest on an intention to treat basis. In multi-item scales with missing data, scale values were calculated if at least half of the items were filled out by imputing the mean of the remaining items. We compared the intervention groups at baseline using χt T2 see . To evalt T2 see . For cont T2 see . The basBecause randomisation took place by family, and because family members were not independent on the outcome measures, statistical significance will be inflated when all women are treated as independent units. At baseline, we included 33 families with multiple participating members (range 2–5), with a total of 80 women. The subsample of mutation carriers consisted of 10 families with multiple members (range 2–3), with a total of 23 women. To counter inflation, we further examined significant effects by incorporating only the first included family member in the analyses.In a previous study of ours, we found that after blood sampling for genetic testing, women affected with breast or ovarian cancer reported a worse well-being and a lower preference for participation in decision-making than women without a previous history of cancer (before the test result (DA_b). Mutation carriers from the control group subsequently received the DA after the test result (DA_a). Of the 44 women from the DA_a group, two (5%) withdrew because of high emotional distress caused by the test result; it is unclear whether their withdrawal was also related to the DA. Of the remaining 42 women from the DA_a group, all viewed the DA. At T3, none of the women withdrew.The participant flow is presented in The DA was viewed once by 49%, and twice or more by 51%. Most respondents (82%) found that it contained exactly enough information, and 13% stated that it contained slightly too little information. In all, 56% reported no negative, 31% a scarcely negative, 12% a rather negative, and 1% a negative emotional reaction towards the information provided. The evaluation forms contained predominantly positive remarks.No significant differences were found between the DA and control groups for anxiety, depression, cancer related distress, and general health (At baseline (T1), no significant differences were found between the DA and control groups . At T2, For strength of treatment preference, decision uncertainty, and preference for decision-making , no signFor subjective knowledge, amount of received information, and satisfaction with quality of information , signifiThe mean risk estimates, the percentage under-, accurate, and overestimates, are presented in For well-being, decision and information related outcomes (P=0.13), the valuation of prophylactic oophorectomy (P=0.15), and risk accuracy of BRCA1/2 related ovarian cancer risk (P=0.18).Significant differences found above were further tested by including only the first family member. Differences between the DA and control groups were no longer significant for the valuation of prophylactic mastectomy (P<0.01) and breast cancer screening (P<0.01). Women affected with cancer valued prophylactic mastectomy higher and breast cancer screening lower in the DA_a group compared to the DA_b group, whereas for women without a previous history this was in the opposite direction.No interactions were found between the DA and history of cancer. No interactions were found between timing of the DA and history of cancer, except for valuation of prophylactic mastectomy . This was beyond the scope of our study, which focused on mutation carriers subsequently. Second, the subsample of mutation carriers is relatively small reducing the power to detect meaningful differences between the women who received the DA either before or after the test result. Third, all treatment choices, even those obtained 2 weeks after a positive test result, are merely intentional as prophylactic surgery is usually postponed for several months until all specialists have been consulted or even longer in young women. However, additional clinical follow-up showed that intended treatment choices, obtained shortly after a positive test result (T3), are strongly predictive of the executed treatment at 9 months after a positive test result: for example, 53% of the women opting for prophylactic mastectomy had undergone this treatment, compared to none of the women opting for breast cancer screening. Fourth, long-term follow-up on the effects of the DA was not obtained because all mutation carriers eventually had received the DA. In the second part of our study, evaluating another DA, including trade-offs and a formal treatment advice derived from a decision model (The strength of our study is that we included a consecutive sample of women covering the eastern part of the Netherlands. Some limitations should be considered. First, we did not evaluate the impact of timing of the DA in women receiving an inconclusive test result (