T1 and GSTM1 genes, and the rs1695 polymorphism in GSTP1 , and risk of breast cancer-specific (n=45) and all-cause (n=99) mortality in a multiethnic, prospective cohort of 533 women diagnosed with stage I-IIIA breast cancer in 1995–1999, enrolled in the Health, Eating, Activity, and Lifestyle (HEAL) Study.Here we assessed associations between null mutations in glutathione-S-transferase (GST)GSTT1 and GSTM1, and the rs1695 polymorphism in GSTP1 by polymerase chain reaction. We assessed associations between breast-cancer specific and all-cause mortality using Cox proportional hazards models.We measured the presence of the null mutation in GSTT1 null , and African American women were more likely to be GSTM1 null . Neither GSTM1 nor GSTT1 null mutations were associated with breast cancer-specific or all-cause mortality. In a model adjusted for body mass index, race/ethnicity, tumor stage and treatment received at diagnosis, the variant Val allele of rs1695 was associated with increased risk of all-cause , but not breast cancer-specific mortality. The GSTT1 null mutation was associated with significantly higher levels of C-reactive protein.Participants with ER-negative tumors were more likely to be GSTM1 and GSTT1 null genotypes had no effect on outcome; however the variant allele of rs1695 appears to confer increased risk for all-cause mortality in breast-cancer survivors.Given the limited sample size of most studies examining associations between GST polymorphisms with breast cancer survival, and the lack of women undergoing more contemporary treatment protocols (treated prior to 1999), it may be helpful to re-examine this issue among larger samples of women diagnosed after the late 1990s, who all received some form of chemotherapy or radiotherapy. GSTM1, GSTT1, and GSTP1 genes, respectively; and these 3 genes have been studied in association with genetic susceptibility to cancer (Strange & Fryer The Glutathione-S Transferases (GST) are a phase II superfamily of cytosolic, mitochondrial and microsomal enzymes that catalyze the conjugation of reduced glutathione to electrophilic centers on a variety of substrates , are associated with a lack of enzyme function and increased vulnerability to cytogenetic damage . The variant allele is associated with lower substrate-specific catalytic activity, including towards the alkylating anticancer agent chlorambucil . Four of the six  and survival. One other large study of 1034 women from Shanghai, China, all treated with adjuvant chemotherapy, found a reduction in risk with the variant GSTP1 Val allele but no association with either GSTT1 or GSTM1 and risk of death  polymorphism  results in an Ile→Val change at codon 105 , reported an association between GSTM1 and GSTT1, the Ile105Val polymorphism in GSTP1), and all-cause and breast-cancer specific mortality in a multi-ethnic cohort of breast cancer survivors drawn from population-based cancer registries. This sample of breast cancer patients diagnosed from 1995–1999 includes a larger number of women undergoing chemotherapy and/or radiotherapy than most prior studies  registries in New Mexico (NM), Los Angeles County (CA), and western Washington (WA). Baseline surveys were conducted on average 6-months post-diagnosis. In NM, we recruited 615 women, ≥18 years, diagnosed with in situ) disease, and 24 women with non-fatal breast cancer events <9 months before their 24-month interview dates to avoid potential confounding from possible recent treatment. The final sample size is 533.944 women completed in-person interviews approximately 30-months following their first interview; 726 women were genotyped; we excluded 169 women with a diagnosis of Stage 0  was used to differentiate GSTP1 genotypes ; non-Hispanic whites (WA); Hispanics; and African Americans.Standardized questionnaire information including medical history, demographic and lifestyle information, was collected at approximately 6- and 30-months post-diagnosis. With participants wearing light indoor clothing and no shoes, weight was measured to the nearest 0.1 kg, and height to the nearest 0.1 cm. All measurements were performed twice, and averaged. Body mass index (BMI) was calculated as kg/mSerum levels of C-reactive protein (CRP) were measured as described previously , Stage I  or Stage II-IIIA  breast cancer based on AJCC stage of disease classification contained within SEER. This analysis includes only women with Stage I-IIIa at diagnosis because few deaths occurred in women with Stage 0 disease. Estrogen receptor (ER) status was categorized as positive, negative, or unknown/borderline. Treatment and additional clinical data were obtained from medical record reviews. Treatment was categorized into 3 groups: surgery only, surgery plus radiation, or surgery with any chemotherapy with or without radiation.Participants were classified as having Stage 0 . Breast cancer-specific mortality was defined as death from breast cancer or end of follow-up, with the same intervals as for all-cause mortality.GSTP1 variant allele were few, heterozygous and homozygous variant allele groups were combined (recessive model). Hazard ratios (HR) and 95% confidence intervals (CI) for breast cancer-specific or all-cause mortality were based on the partial likelihood for Cox's proportional hazards model . Differences in distributions of categorical variables were compared using the Chi-square test. As the numbers of patients homozygous for the odel Cox . The pro2; ≥18.5 and <25 kg/m2; ≥25 and <40 kg/m2; ≥40 kg/m2); SEER summary tumor stage  and treatment received at diagnosis . Covariates considered but not included in the final model (they did not significantly change the likelihood ratio score): menopausal status, education, smoking status, tamoxifen use, and ER status. The Wald statistic was used to test for trend across levels.We based variable inclusion on a likelihood ratio test, with the following covariates included in models: race/ethnicity/study-site (to adjust for different distributions of race/ethnicity by study site); BMI .GSTM1 null mutation, and significantly more African American women carried it compared to NHW or Hispanics (χ2=34.36 P<0.0001). 79.6% of participants carried at least one GSTT1 null mutation; there was no difference in the proportion of carriers and non-carriers across racial/ethnic groups. For the GSTP1 Ile105Val polymorphism, 58.2% of participants carried at least one variant allele; there were no differences across racial/ethnic groups. The GSTP1105Ile/Val polymorphism was in Hardy-Weinberg equilibrium (P>0.05). Median follow-up time was 11.29 years.Mean age of participants was 57.6 years Table ; more thGSTM1 null mutation were more likely to have stage 3 tumors (χ2=3.87 P=0.05) compared to non-carriers . There were no significant associations between the GSTP1 polymorphism and patient characteristics. When we excluded participants who had surgery only from the analysis, these associations were no longer significant (P=0.06 for both) but this may be due to smaller numbers.When we compared distribution of genotypes by participants’ characteristics , carriers of the GSTT1 null mutation had significantly higher levels of CRP compared to non-carriers . Levels were also significantly higher in participants with diabetes , but there were no differences in participants without diabetes. In contrast the GSTM1 null mutation was associated with significantly lower levels of CRP among diabetics only (mean: 4.83 vs. 14.61 mg/dL). There were no associations between GSTP1 and CRP (data not shown).Carriers of the GSTM1 or GSTT1 null mutations, and breast- or all-cause mortality. Heterozygous carriers of the GSTP1 Ile105Val polymorphism had a significantly increased HR for all-cause mortality, compared to wild-type homozygotes . When examined as a recessive model , the variant allele was associated with an increased risk of all-cause mortality . There was no association between the GSTP1 Ile105Val polymorphism and breast cancer-specific mortality. When we examined risk in women who were postmenopausal at baseline (N=341), HR was similar to that in the entire population for all 3 polymorphisms (data not shown). Finally, we examined associations with mortality in patients who received any treatment (chemotherapy/radiotherapy): there were no differences in associations for GSTT1 and GSTM1. The association for GSTP1 for homozygous Val/Val changed from HR=1.24 95% CI 0.59-2.58) to HR=0.60 95% CI 0.20-1.75 , but none reached statistical significance (data not shown). Breast cancer-specific mortality demonstrated a different pattern, with carriers of GSTT1 null mutation/GSTM1 present/GSTP1105Ile/Ile genotypes associated with a reduced risk  compared to participants with GSTM1 present/GSTT1 present/homozygous wild-type for GSTP1. However this association was not significant (P=0.06).We performed a 3-way gene analysis, examining combinations of GSTP1, GSTM1 and GSTT1 polymorphisms in patient subgroups, using fully adjusted models. The GSTP1 Val allele was associated with an approximate 2-fold increased risk of all-cause mortality in patients with ER-positive tumors, compared to those with ER-negative, though this was not significant (P=0.08). There was no evidence of effect modification for other subgroups examined. Due to limited power we did not examine associations for breast cancer-specific mortality in these subgroups.We next analyzed the same endpoints for GSTP1).There are few studies on the role of GST isoenzymes on mortality in breast-cancer survivors drawn from community practice. As described earlier, the majority of these studies had small sample sizes, were based on participants diagnosed prior to 1999, and on women undergoing chemotherapy and/or radiotherapy. In addition, most examined only one GST gene  was associated with increased risk of all-cause mortality, but not breast cancer-specific mortality, in a cohort of 533 breast-cancer survivors. The amino acid substitution in GSTP1 Ile105Val, results in an enzyme with altered activity, , homozygous carriers of the GSTM1 null mutation in patients with diabetes only; and higher levels of CRP in carriers of the GSTT1 null mutation; the latter differed from results in patients with diabetes and smokers (lower CRP among carriers of the GSTT1 null mutation); , suggestions for further study include examining larger studies of pooled data of women diagnosed at similar time-periods who underwent similar treatment regimens, thus enhancing power to detect associations between GST isoenzymes and longer-term survival.


Pregnancy-associated breast cancer refers to breast cancer diagnosed during pregnancy, lactation, or within twelve months postpartum. Recent studies suggest that, when matched for age and stage, the prognosis of pregnancy-associated breast cancer is comparable to non-pregnancy-associated breast cancer. However, the risk for breast cancer recurrence associated with subsequent pregnancies in this population is not clear.We describe the case of a Caucasian woman who was initially treated for pregnancy-associated breast cancer at age 23, three months after the birth of her third child. She underwent a total mastectomy with axillary node dissection, followed by chemotherapy and hormonal therapy. Ten years later, when the patient was 24 weeks pregnant with her fourth child, she presented with an ipsilateral chest wall recurrence of breast cancer. To the best of our knowledge, this represents the first reported case of a pregnancy-associated recurrence in a patient previously treated for pregnancy-associated breast cancer.The case described here is the first report of a second occurrence of pregnancy-associated breast cancer. This case raises the possibility that pregnancy may represent a unique trigger for breast malignancy in a specific cohort of women. Although there is data showing no increase in the risk of recurrence for women who become pregnant after breast cancer treatment, pregnancy-associated breast cancer may be a distinct clinical category where subsequent pregnancies after treatment may confer an increased risk of recurrent disease. In 10 % of women diagnosed with breast cancer (BC) under the age of 40 years, the disease is associated with pregnancy . With moHistorically, it was assumed that PABC carried a worse prognosis than BC . PABC usWith women delaying childbirth, there is a larger group of premenopausal survivors of BC interested in subsequent childbearing . In the In the following case, a patient who had been treated for PABC developed another episode of PABC during a subsequent pregnancy ten years after her initial diagnosis. Our case, coupled with the current theories regarding pregnancy and breast malignancy, strongly suggests that women with PABC may represent a specific subgroup of women for whom a subsequent pregnancy may impart unique risks. This observation supports the need to consider a prior history of PABC when counseling BC survivors concerning the potential risks of future pregnancies.BRCA1 and BRCA2 mutations. At the time of her initial presentation, an ultrasound of her left breast revealed a 1.7 cm mass in the area of palpable abnormality, and ultrasound guided fine needle aspiration (FNA) was suspicious for carcinoma. Extensive microcalcifications in her left breast were demonstrated on mammography. A left breast lumpectomy showed extensive intraductal and invasive ductal carcinoma with multifocally involved margins. The largest single focus of invasive cancer measured 1.5 cm. The cancer was 38 % estrogen receptor positive and 35 % progesterone receptor positive. Our patient then underwent a left total mastectomy with axillary lymph node dissection and transverse rectus abdominis myocutaneous flap reconstruction. The pathology at the time revealed additional extensive intraductal carcinoma, and axillary nodes showed no evidence of disease. Postoperatively, she received chemotherapy with cyclophosphamide, doxorubicin (Adriamycin) and 5-fluorouracil and was maintained on tamoxifen and leuprolide for three years. She was monitored closely and was without evidence of recurrence for approximately 10 years.A 23-year-old Caucasian woman first presented to our department 15 years ago with a palpable mass in her left breast three months after the delivery of her third child. At the time, she was breast-feeding. She had no family history of breast or ovarian cancer, and no personal history of breast disease. Menarche occurred at age thirteen, and she was nineteen when she delivered her first child. In view of her young age and Ashkenazi Jewish heritage, she subsequently underwent genetic testing, which was negative for Five years ago, our patient was evaluated with mammography, ultrasound and magnetic resonance imaging, which showed no evidence of malignancy in the contralateral breast. Our patient returned six months later, 20 weeks pregnant, at which time her clinical exam was unremarkable. One month later, our patient presented in the 24th week of pregnancy with a palpable mass in the upper-outer portion of her reconstructed left breast. On examination, a subcutaneous mass was appreciated, associated with the native skin of her reconstructed left breast. FNA of the mass identified malignant cells. Our patient underwent wide local excision of the mass. Histology confirmed poorly differentiated breast carcinoma invading into adipose tissue. The tumor, measuring 0.4 cm in its widest dimension, was noted to be estrogen receptor negative, but highly progesterone receptor positive (95 %) with overexpression of human epidermal growth factor receptor 2/Neu. After delivery of a healthy male child at term, our patient was treated with tamoxifen and trastuzumab and underwent radiation therapy to her left chest wall. At present, she remains on tamoxifen and is doing well without recurrent disease.Historically, women were counseled against pregnancy after BC treatment owing to concerns regarding increasing risk for recurrence and mortality. Our increased understanding of the hormone responsiveness of BC produced the theory that the elevated levels of estrogen and progesterone in pregnancy could increase the risk of breast cancer recurrence. However, contemporary data shows no increased risk of recurrence and no difference in survival in women who had children after BC therapy when compared with women who chose not to become pregnant ,10,11. AAt this point, it is possible that there are insufficient numbers of survivors of PABC who have subsequently become pregnant to establish any associated increase in risk for recurrence. This case report illustrates a recurrence of PABC that, to the best of our knowledge, has not previously been described in the medical literature. In this case, the time lag of 10 years indicates that the second episode of PABC is likely to be a second primary. This is supported by the slightly different biomarker profile. In our patient, who lacks an identified genetic predisposition to BC, the unique hormonal milieu of pregnancy may have been a specific trigger for malignancy.In the modern era, we have come to recognize that women can bear children after BC treatment without compromising their survival. There is insufficient data to assess the impact of additional pregnancies on the risk of recurrence and survival for women with PABC. The case in this report indicates that PABC survivors, unlike BC survivors, may be at risk for recurrent disease during subsequent pregnancies. They may be uniquely susceptible to the hormonal milieu of pregnancy, producing an increased risk for recurrence with subsequent pregnancies. In the absence of specific information regarding this subgroup of women, caution should be exercised when counseling survivors of PABC regarding subsequent pregnancy. With this understanding, we suggest that women who choose to become pregnant after PABC should be monitored more closely during pregnancy and postpartum. Longitudinal studies will be important to establish the BC-related outcomes of women who become pregnant after treatment of PABC. Further research is needed to elucidate the underlying mechanisms and specific risks, and allow for appropriate evidence-based counseling of survivors of PABC.Written informed consent was obtained from the patient for publication of this case report. A copy of the written consent is available for review by the Editor-in-Chief of this journal.BC, Breast cancer; FNA, Fine needle aspiration; PABC, Pregnancy-associated breast cancer.The authors declare that they have no competing interests.FS is the patient’s surgeon and was a major contributor in writing and editing the manuscript. JB and AC performed the medical literature review on the topic of PABCs and were also major contributors in writing the manuscript. JC reviewed the statistics and was a major contributor in designing, writing and editing the manuscript. All authors read and approved the final manuscript.


The overall prevalence of a hereditary breast cancer phenotype was 6.2% (95%CI: 5.67-6.65). These findings identified a problem of significant magnitude in the region and indicate that genetic cancer risk evaluation should be undertaken in a considerable proportion of the women from this community. The large proportion of women who attended GCRA (72.3%) indicates that the program was well-accepted by the community, regardless of the potential cultural, economic and social barriers. In 2004, a population-based cohort (the Núcleo Mama Porto Alegre - NMPOA Cohort) was started in Porto Alegre, southern Brazil and within that cohort, a hereditary breast cancer study was initiated, aiming to determine the prevalence of hereditary breast cancer phenotypes and evaluate acceptance of a genetic cancer risk assessment (GCRA) program. Women from that cohort who reported a positive family history of cancer were referred to GCRA. Of the 9218 women enrolled, 1286 (13.9%) reported a family history of cancer. Of the 902 women who attended GCRA, 55 (8%) had an estimated lifetime risk of breast cancer ≥ 20% and 214 (23.7%) had pedigrees suggestive of a breast cancer predisposition syndrome; an unexpectedly high number of these fulfilled criteria for Li-Fraumeni- A significant challenge in Brazil and other Latin American countries is the inclusion of new health technologies, including genetic risk assessment and testing, in health care systems that have significant budget constraints.Latin America is formed by low-to-medium income countries with health indicators that are evolving towards patterns seen in developed nations. The extensive ethnic and cultural diversity seen in these countries reflects their history and different degrees of admixture between native and immigrant populations. Brazil is the largest and most populated country in Latin America, with ~170 million inhabitants. Although Brazil has the eighth largest gross national product in the world, the average per capita income ranks only 39Sistema Único de Saúde). One special SUS program designed for community-based health care is the Programa Saúde da Família , created in the mid-90s and based upon multidisciplinary teams composed of a physician, nurse, 1-2 nursing assistants and 4-6 lay community health workers. Each team provides primary health care for a geographically defined group of approximately 600 families. The Family Health program has expanded rapidly and currently provides health care to about half of the population , Brazil's southernmost state, has one of the highest incidences of breast cancer in the country and an increasing breast cancer mortality rate, despite efforts to improve the coverage for breast cancer screening . Breast cancer is also the leading cause of death by cancer in young women (30-49 years) from this region  which accounts for most cases of hereditary breast cancer (HBC) worldwide  (PTEN (associated with Cowden's syndrome)    PTEN (asyndrome)  and CHEKLittle attention has been given to the identification and study of hereditary breast cancer phenotypes in the community and primary health care services  was started in Porto Alegre, the capital of the southern Brazilian state of Rio Grande do Sul. The cohort intends to collect demographic, epidemiologic and risk factor data from a large sample of women 15 years old and test a model for community-based breast cancer screening for women between the ages of 40 and 69 years, as described elsewhere  was obtained from the institutions involved and enrollment in the study required signature of informed consent. Active recruitment was initiated after six months if the patients referred by the Family Health program did not reach NMPOA. In this case, three attempts were made to schedule a visit by telephone, followed by a letter of invitation and a search for the patient by community agents. If all of these strategies failed, or if three scheduled appointments were not kept, no further contact was attempted.To determine the weighted prevalence of a hereditary breast cancer phenotype in the population being studied, two groups of patients were evaluated: (a) 885 unrelated women with a family history of cancer and (b) 910 unrelated women of the same cohort with no family history of cancer upon recruitment at the primary health care unit and who were invited to participate in this study during their annual mammographic screening examinations. Recruitment was done consecutively during a period of 12 months. The evaluation included an interview, an estimation of the risk of breast cancer, and registration of the family history in pedigrees of at least three generations. The presence of criteria for breast cancer predisposition syndromes was assessed by three clinical geneticists  who independently reviewed each pedigree. The group of patients referred for genetic risk evaluation was also the group used for the validation of the seven question instrument for identification of hereditary breast cancer families .BRCA1 or BRCA2 mutation were determined for each patient by using mutation prevalence tables and a modified Couch (Penn II) mutation probability model  and laterally as possible and including paternal lineages. Confirmation of the family history of cancer was attempted in all cases and pathology reports, medical records and/or death certificates were obtained whenever possible (unpublished data). Estimated lifetime risks of breast cancer (ELTR) were obtained by using Claus tables and the Gail and Tyrer-Cuzick models (ty model . All of All of the patients were encouraged to perform monthly breast self-examinations. Women with ELTR < 0.2 (Risk I) were referred back to their primary health care unit for prospective follow-up as determined by their age and other non-genetic risk factors . Patients with ELTR ≥ 0.2 (Risk II) and those who fulfilled criteria for hereditary breast cancer syndrome (Risk III) were referred for clinical breast evaluations at 6-month intervals. In addition, for women in risk categories II and III, a recommendation was made for annual mammography intercalated with breast magnetic resonance imaging at 6-month intervals; breast ultrasound was recommended for those with very dense breasts and/or whenever clinically indicated. Women in risk category III began screening at the age of 25 years or as soon as a risk was identified. Those with an increased risk for tumors other than breast cancer were referred to tertiary care centers for inclusion in comprehensive cancer screening programs. Those with criteria for a breast cancer predisposition syndrome were offered genetic testing and the results of this investigation will be described elsewhere.After conclusion of the genetic evaluation, written reports and a brochure with key information about the prevalence, treatment and prevention of breast cancer, as well as information about hereditary breast cancer, were mailed to all of the patients. The vast majority of the women included in this study relied exclusively upon the public health care system. Although mammographic screening and clinical follow-up are available in this system outside the cohort study described here, access to these facilities is limited and most women would not have been able to comply with the recommended screening guidelines if they had not been included in this study. Magnetic resonance imaging of the breast is not routinely available in the public health care system.2 test was used to compare the distribution of the answers to the family history questionnaire between attenders and non-attenders, whereas ANOVA was used to compare the mean values of the risk estimates and the history of breast cancer among the different risk categories. Kappa coefficients were used to assess the agreement between results provided by the instrument and the estimated genetic cancer risk. The weighted prevalence of hereditary breast cancer phenotypes was calculated using the following weighting: group with a family history of cancer (885/1285) and group with no family history of cancer (910/7933). All data handling and statistical analyses were done using the statistical software package SPSS (version 14.0), and a value of p < 0.05 indicated significance.Quantitative variables were expressed as the mean ± SD, whereas categorical variables  were recorded as absolute and/or relative frequencies. The χThe women enrolled in this study were residents of seven regions in Porto Alegre. These regions correspond to 64% of the total area of the city and ~35% of its population. All women relied almost exclusively on the Family Health Program for health care. Although specific demographic data for this population of the Family Health Program clients was unavailable, data from the most recent municipal census indicate that 49.2% of the inhabitants in these regions have < 8 years of education (5.6% are illiterate), 20% live in temporary homes and 6.5% of family providers have no income .Of the 9218 women enrolled in the cohort, 1286 (13.9%) answered positively to at least one of the seven questions about a family history of cancer  and thosand ovarian cancer in a relative, for which non-attenders provided a greater number of positive answers  remained in most cases (Kappa coefficient = 0.069).The demographic data of women with a positive family history who underwent GCRA (attenders) differed significantly in some aspects from those of women who did not undergo GCRA (non-attenders). Non-attenders were generally younger, less educated, and had undergone a breast biopsy less often than attenders (p < 0.001). For most of the questions in the family history questionnaire, there was no significant difference between the proportion of positive answers between attenders and non-attenders. Exceptions to this included: (a) the question about a family history of breast and/or ovarian cancer in ≥ 2 relatives, for which attenders provided a greater number of positive answers, and (b) the question about the combined occurrence of breast  answers . The latlike syndrome  was similar to that described in other population-based samples, although only a few studies have addressed this issue to date  may be more frequent in the population than previously estimated. In addition, several recent reports have indicated that a deleterious germline mutation in TP53, R337H, might be very prevalent and related to a founder effect in southern Brazil  and may have little sensitivity for identifying germline TP53 mutations in other countries, the unexpectedly high number of criteria-positive patients suggests that LFL may indeed be a common phenotype for hereditary breast cancer in and around Porto Alegre. Future results from germline mutation testing should help to clarify this issue and contribute to our understanding of the applicability and discriminatory capacity of breast cancer risk estimation models, current diagnostic criteria and mutation prediction models for hereditary breast cancer syndromes in this population. If such a high prevalence of LFS/LFL syndrome is confirmed among families in this geographic area, an effort will have to be made to facilitate the identification of high-risk individuals and delineate effective cancer screening and prevention programs in these subjects.Already ten years ago  in the U.S.A., the outcome of genetics referrals was evaluated in a group of women with estimated BRCA mutation probabilities ≥ 10%. Within six-months after referral, 36% of the patients had undergone genetic evaluations (acceptors), 27% still intended to seek (intenders) and 36% refused such assessment (decliners). Population-based mammographic screening programs worldwide have also reported compliance rates of 61%-83% (The relatively large proportion of women (72.3%) who attended the proposed GCRA was very encouraging and comparable to that encountered in similar studies, In this study, the number of attenders also has to be interpreted in light of the difficulties that most of these women face to seek advice, health care and cancer prevention opportunities. First, there might be a cultural difference in cancer-associated risk perception and/or a difficulty in understanding the impact of preventive interventions. The way of dealing with risk is not only influenced by risk perception but also by culture. This was clearly demonstrated by i.e., facilitate transportation, nutrition and childcare) in the difficult task of maintaining compliance to the recommended guidelines. A more detailed study of non-attenders may provide better clues on how to improve coverage for programs such as these, thereby ensuring that most high-risk patients have access to the information and preventive interventions they require.For all of these reasons, programs such as that described here, which attempt to identify and prospectively follow women with an increased risk of cancer, need to consider the importance of patient education and social interventions (In conclusion, the implementation of a GCRA program for an underserved community in southern Brazil revealed that the overall prevalence of a hereditary breast cancer phenotype in this population-based sample of women was 6.2%, which may be a matter of considerable concern in this region. In addition, the establishment of breast cancer risk categories allowed the identification of higher risk women who may benefit from more intensive screening. The relatively high number of women who attended GCRA sessions after an initial referral suggested that the program was well accepted by the community and is feasible, regardless of potential cultural, economic and social barriers.


Prospective epidemiologic studies have consistently shown that levels of circulating androgens in postmenopausal women are positively associated with breast cancer risk. However, data in premenopausal women are limited.in situ) and 683 individually-matched controls were included. Matching variables included age and date, phase, and day of menstrual cycle at blood donation. Testosterone, androstenedione, dehydroandrosterone sulfate (DHEAS) and sex hormone-binding globulin (SHBG) were measured using direct immunoassays. Free testosterone was calculated.A case-control study nested within the New York University Women's Health Study was conducted. A total of 356 cases (276 invasive and 80 Ptrend = 0.04), and for free testosterone were 1.0 (reference), 1.2 , 1.5 , 1.5 , and 1.8 . A marginally significant positive association was observed with androstenedione (P = 0.07), but no association with DHEAS or SHBG. Results were consistent in analyses stratified by tumor type , estrogen receptor status, age at blood donation, and menopausal status at diagnosis. Intra-class correlation coefficients for samples collected from 0.8 to 5.3 years apart (median 2 years) in 138 cases and 268 controls were greater than 0.7 for all biomarkers except for androstenedione (0.57 in controls).Premenopausal serum testosterone and free testosterone concentrations were positively associated with breast cancer risk. In models adjusted for known risk factors of breast cancer, the odds ratios for increasing quintiles of testosterone were 1.0 (reference), 1.5 , 0.9 to 2.3), 1.2 , 1.4  and 1.8 . Results from other prospective studies are consistent with our results. The impact of including testosterone or free testosterone in breast cancer risk prediction models for women between the ages of 40 and 50 years should be assessed. Improving risk prediction models for this age group could help decision making regarding both screening and chemoprevention of breast cancer. Prospective epidemiologic studies have consistently shown that circulating androgens in postmenopausal women are positively associated with breast cancer risk -8, an asWe report here the results of a case-control study nested within the New York University (NYU) Women's Health Study cohort. Prediagnostic concentrations of testosterone, androstenedione, dehydroepiandrosterone sulfate (DHEAS) and sex hormone-binding globulin (SHBG) were measured in serum samples from 356 incident cases and 683 controls who were premenopausal at enrollment , and the association of these biomarkers with subsequent risk of breast cancer was assessed.The NYU Women's Health Study (NYUWHS) enrolled 14,274 women 34 to 65 years old at a breast cancer screening center in New York City between 1985 and 1991 . Women wWomen were classified as premenopausal at enrollment/first blood donation if they reported at least one menstrual cycle in the six months prior to their visit. To determine the phase of cycle at blood donation, the start date of the menstrual period prior to the visit was recorded, and women were asked to return a postcard indicating the start date of their next menstrual cycle. Seventy-two percent of the women returned the postcard. For women with length of cycle 20 to 41 days, phase of cycle was calculated based on the number of days between the date of blood donation and the start date of the next menses: < 12 days: luteal; 12 to 16 days: peri-ovulatory; 17 to 19 days: late follicular; ≥20 days: early follicular. For the 15% of women who did not return the postcard but reported regular menstrual cycles, the date of menstrual period prior to the visit and the usual length of cycle were used to compute the phase of cycle at blood donation. Women who had had a hysterectomy without total oophorectomy prior to natural menopause and were less than 52 years of age at enrollment were also classified as premenopausal, with subsequent verification of menopausal status by follicle-stimulating hormone (FSH) measurements on a nested case-control basis. Women with a concentration < 12.75 mIU/mL were confirmed as premenopausal. For these women (7%), the phase of cycle at blood donation was coded as unknown. Phase of cycle was also coded as unknown for women who did not return the postcard and reported having irregular cycles and for women who returned the postcard but had a length of cycle less than 20 or more than 41 days (13%). The study was approved by the Institutional Review Board at the New York University School of Medicine.Breast cancer cases were identified through active follow-up of the cohort by mailed questionnaires approximately every two to four years and telephone interviews for non-respondents, as well as record linkages with state cancer registries in New York, New Jersey, and Florida, and with the US National Death Index. A capture-recapture analysis estimated the breast cancer case ascertainment rate in our cohort to be 95% . Only inFor each case, two controls were selected at random from the appropriate risk set. The risk set for a case consisted of all women premenopausal at enrollment who were alive and free of cancer at the time of diagnosis of the case (index date) and who matched the case on age at enrollment/first blood donation (± 6 months), date of enrollment (± 3 months), number  and dates (± 6 months) of subsequent blood donations, and phase  and day of menstrual cycle at the first blood donation.All assays were conducted in the Hormones and Cancer Group at the International Agency for Research on Cancer in Lyon, France. Assays were selected based on the results of a validity study . Testostth and 90th percentiles were calculated for the hormonal measurements and a mixed-effects regression model accounting for the matching was used to compare concentrations in cases and controls. Odds ratios were estimated using conditional logistic regression. Biomarkers were analyzed as quintiles based on the distribution of cases and controls combined, and trend tests were carried out using ordered categorical variables. We also conducted an analysis using the mean hormone level for women who had two samples, and the single available measurement for the remaining women. Finally, each biomarker was also analyzed after log2-transformation to estimate odds ratios corresponding to a doubling in concentration and to compute a trend test on the continuous scale. Adjusted models included known risk factors for breast cancer, that is, age at menarche, family history of breast cancer, parity, age at first birth, history of breast biopsy, and body mass index. Analyses were also done stratifying by tumor type (invasive and in situ), estrogen receptor status, age at enrollment, menopausal status at index date, lag time between blood donation and diagnosis of the case, and in the subgroup of women who had reported a history of regular menstrual cycles as well as five to seven cycles in the six months prior to blood donation. Analyses stratified by body mass index (BMI) and menopausal status at diagnosis and in the subgroup of women with five to seven regular menstrual cycles in the six months prior to blood donation were carried out using unconditional logistic regression, controlling for the matching factors, to avoid exclusion from the analysis of matched sets whose case and controls would be in different strata. All stratified analyses were carried out on log2-transformed hormones because of the smaller sample sizes in subgroups. The P-value obtained when adding a cross-product term to the model containing main effects was used to assess interaction. All statistical tests were two-sided.Case and control characteristics were compared using the conditional logistic regression model, to take into account the individual matching. Median, 10The intra-class correlation coefficient (ICC)  was usedin situ) were diagnosed by 1 January 2000. Ten cases were excluded because of FSH concentrations compatible with postmenopausal status. As a result, 356 cases (276 invasive and 80 in situ) are included in this analysis. Among the initially selected individually-matched controls, 29 were excluded because of their FSH concentrations, resulting in the inclusion of 683 controls in this analysis. For 138 cases and 268 controls, serum samples collected at two separate visits were analyzed.Among the 7,220 (50.6%) women who were premenopausal at the time of initial blood donation, 366 breast cancer cases (285 invasive and 81 P = 0.04) and cases tended to have a later age at first full-term pregnancy than controls (P = 0.03). There was also a higher proportion of women with a family history of breast cancer among cases than controls (P = 0.004). There was no difference between cases and controls in the proportions of overweight  and obese  women. Seventy-six percent of the cases and 78% of the controls had a history of regular cycles and also reported five to seven cycles in the six months prior to enrollment.Table P = 0.01) as were free testosterone concentrations . The median androstenedione concentration was marginally higher in cases than controls , whereas no statistically significant difference was observed for DHEAS and SHBG.As was observed in longitudinal studies of premenopausal women , the higP = 0.04) and free testosterone (P = 0.01), with odds ratio (95% CI) of 1.8  and 1.8 , respectively, for women in the highest versus lowest quintile. The increase in risk appeared more linear for free testosterone  than testosterone . A marginally significant trend (P = 0.07) of increasing risk with increasing concentration of androstenedione was also observed, while no significant association was observed between concentrations of DHEAS and SHBG with breast cancer risk. Associations were similar in analyses using the mean hormone level for women who had two samples and the single available measurement for the remaining women, except for testosterone for which higher odds ratios were observed when the average was used if available.Table in situ tumors, as well as for tumors diagnosed before and after menopause. Odds ratios greater than one were also observed for estrogen receptor-negative tumors, although the associations were weaker than for estrogen receptor-positive tumors and not statistically significant.Table Table We observed positive associations between premenopausal concentrations of total and free testosterone and breast cancer risk, with women in the highest quintile having a risk approximately 80% greater than women in the lowest quintile. We also observed a marginally significant positive association with androstenedione but no association with DHEAS or SHBG. The observation of similar associations (except for testosterone for which the odds ratios increased slightly) in an analysis using the mean hormone level for women who had two samples and the single available measurement for the remaining women strengthened our conclusions. There was no evidence of heterogeneity in the associations of total and free testosterone with breast cancer risk in subgroups according to tumor type, estrogen receptor status, age and BMI at enrollment, menopausal status at diagnosis and lag time between blood donation and diagnosis.Two prospective studies also reported statistically significant positive associations between premenopausal concentrations of testosterone and free testosterone and breast cancer risk ,17, threIt has been suggested that including circulating concentrations of sex hormones could improve risk prediction models ,39. In fet al. that adjustment for concentrations of estradiol in premenopausal women did not affect risk estimates associated with testosterone concentrations [Positive associations between circulating androgens in postmenopausal women and risk of breast cancer have been observed consistently , althougtrations . It is oIn addition to their role as estrogen precursors, it has been proposed that androgens directly impact cell proliferation ,55, possAndrogens, in particular testosterone, have been proposed for relief of menopausal symptoms, in particular sexual desire deficit . As sympThe NYUWHS was designed primarily to examine the association of endogenous sex hormones with risk of breast cancer. We therefore excluded women taking exogenous estrogens and collected data on date of next menstrual period which allowed us to calculate the phase of cycle more precisely than some other studies. Other strengths of our study include the large number of cases, which allowed us to examine various subgroups and the availability of two serum samples in a fairly large number of both cases and controls which allowed us to show that a single androgen concentration measurement is quite representative of a woman's concentration over several years. This is despite the fact that we did not control for time of day of blood donation, and, therefore, for possible circadian variations of androgen production. A weakness of our study is that we used radioimmunassays without an extraction step, and the sensitivity and specificity of such assays have been questioned ,62. It sIn conclusion, and in agreement with other cohorts, we observed associations between pre-diagnostic concentrations of total and free testosterone in premenopausal women and risk of breast cancer. These results suggest that androgen concentrations should be considered for inclusion in risk prediction models for women between the ages of 40 and 50, which could help in decision making regarding both screening and chemoprevention of breast cancer.BMI: body mass index; CI: confidence interval; DHEAS: dehydroandrosterone sulfate; EPIC: European Prospective Investigation into Cancer and Nutrition; ER: estrogen receptor; FSH: follicle-stimulating hormone; ICC: intraclass correlation coefficient; NYUWHS: New York University Women's Health Study; SHBG: sex hormone-binding globulin.The authors declare that they have no competing interests.AZJ participated in the design of the study, statistical analysis, and manuscript preparation. YA performed the statistical analysis and contributed to manuscript preparation. RK participated in the conception of the study and manuscript preparation. SR performed serum hormone analyses and contributed to manuscript preparation. SS contributed to manuscript preparation. ML participated in the statistical analysis and manuscript preparation. AAA participated in data acquisition and manuscript preparation. PT and RES participated in the conception and design of the study and manuscript preparation. KLK participated in the design of the study, the statistical analysis and manuscript preparation. All authors read and approved the final manuscript.


Risk factors for breast cancer can be allocated to one of four groups:1 Family History/genetic2 Reproductive/hormonal3 Proliferative benign breast disease4 Mammographic densityThese four factors have now been thoroughly studied and accurate quantitative estimates for the risk are now available for many of them. The most useful summary comes from the Oxford collaboration, which has now produced a series of papers estimating risk for individual factors. Less is known about the possible interaction between these factors and virtually nothing is known about how different factors influence the risk of different types of breast cancer e.g. oestrogen receptor positive versus negative tumours. Risk factors appear to be largely independent and this facilitates building a model to predict risk for individuals. Previous models have focused on either non-genetic factors  where imHere we briefly review the main risk factors for breast cancer and describe our own model  and a co


The objective of this study was to estimate the risk of contralateral breast cancer in BRCA1 and BRCA2 carriers; and measure the extent to which host, family history, and cancer treatment-related factors modify the risk.Patients were 810 women, with stage I or II breast cancer, for whom a BRCA1 or BRCA2 mutation had been identified in the family. Patients were followed from the initial diagnosis of cancer until contralateral mastectomy, contralateral breast cancer, death, or last follow-up.vs 16.8% P=0.003). Women aged <50 years with two or more first-degree relatives with early-onset breast cancer were at high risk of contralateral breast cancer, compared with women with fewer, or no first-degree relatives with breast cancer (50 vs 36% P=0.005). The risk of contralateral breast cancer was reduced with oophorectomy .Overall, 149 subjects (18.4%) developed a contralateral breast cancer. The 15-year actuarial risk of contralateral breast cancer was 36.1% for women with a BRCA1 mutation and was 28.5% for women with a BRCA2 mutation. Women younger than 50 years of age at the time of breast cancer diagnosis were significantly more likely to develop a contralateral breast cancer at 15 years, compared with those older than 50 years (37.6 The risk of contralateral breast cancer risk in BRCA mutation carriers declines with the age of diagnosis and increases with the number of first-degree relatives affected with breast cancer. Oophorectomy reduces the risk of contralateral breast cancer in young women with a BRCA mutation. BRCA1 or the BRCA2 gene face a high lifetime risk of breast cancer  or those who resided outside of North America were excluded. It was not necessary to be a proven carrier of the mutation found in the family to be included in the study; however, affected women who were known to be non-carriers were excluded. All study procedures were approved by the Institutional Review Boards at each of the participating centres.We identified a total of 1866 breast cancer cases in 615 families. Of the total 1866 cases of breast cancer, 417 women were excluded because the date of diagnosis indicated on the pedigree was before 1975, and 70 women were excluded because the age of diagnosis was above 65 years. An additional 29 women were known not to carry the familial mutation and were therefore excluded. In total, 19 women were excluded because they had a diagnosis of other cancer before breast cancer, and 26 women were excluded because they were treated outside of North America.We were able to obtain identifying information for 993 of the remaining 1305 women (76%). An attempt was made to contact each of these or her next-of-kin to obtain permission to review the medical records. Overall, 19 women (or their next of kin) refused to provide consent for the release of the medical records. The medical record was requested from the hospital where treatment was received for the remaining 974 women. In 76 cases, the hospital was not able to locate the record or did not forward the requested documents. The medical record was obtained for the remaining 898 women (92%).After review of the medical records, an additional 52 women were excluded. Of these, 32 women were excluded because tumour stage was greater than two; 18 women were excluded because the tumour was non-invasive (DCIS or LCIS), and 2 women were excluded because they refused treatment. The remaining 846 women were included in the analysis. In summary, of the 1866 breast cancers, 613 were ineligible and we were able to enroll 846 of the remaining 1253 eligible cases (68%). 177 of the 846 women were deceased (20.9%).The medical treatment records and pathology documents were reviewed. We established whether the tumour was unilateral or bilateral at initial diagnosis. We recorded tumour size (in cm), nodal status (positive/negative), and tumour grade (I to III). Where possible, we abstracted information on both mitotic and nuclear grade. Status of ER was recorded as positive, negative, equivocal, or unknown. We recorded the use of chemotherapy (yes/no), tamoxifen (yes/no), and radiotherapy (yes/no). We established whether or not the patient had undergone a bilateral oophorectomy, and if so, the date of the operation. In some cases, the contralateral mastectomy and/or oophorectomy were performed several years after the initial surgical treatment; the dates of these late treatments were recorded. We reviewed the dates of diagnosis of all contralateral breast cancers reported in the cohort. Only invasive contralateral cancers were included. No woman was diagnosed with contralateral breast cancer after she was diagnosed with distal metastases. Information on family history was collected on the patients by review of the pedigree at the time of ascertainment. Each patient was classified according to the number of documented first-degree relatives with early-onset breast cancer (diagnosed at or below age 50 years) as zero, one, or two or more affected relatives.vs BRCA2). We also compared the risk of contralateral breast cancer for subgroups defined by each of the four treatments individually . Hazard ratios were estimated using the Cox proportional hazards model, implemented in SAS . In these analyses, oophorectomy was treated as a time-dependent variable. The hazard ratios were adjusted for age at first cancer diagnosis, mutation status (BRCA1 vs BRCA2), family history, and other treatments received.A survival analysis was performed. We considered the woman to be at risk for contralateral breast cancer from the date of the first surgical procedure to the first of contralateral breast cancer, contralateral mastectomy, death from breast or ovarian cancer, death from another cause, or the date of last follow-up. The date of the last follow-up was defined as date of last contact or death. Survival curves were constructed and compared for subgroups of women defined by age , family history (number of first-degree relatives with breast cancer), and by mutation status  are presented in Patients were diagnosed between 1975 and 2008. The subjects were followed for a mean of 11.1 years (range 0.1–32.9 years). In all, 78.8% of the subjects were alive at the time of last follow-up. In total, 149 subjects (18.4%) were diagnosed with a contralateral breast cancer. All contralateral cancers were confirmed with the medical record. The mean time interval period between the diagnosis of the first breast cancer and the diagnosis of the contralateral breast cancer was 5.7 years (range 0.2–15 years).BRCA1 vs BRCA2), number of affected first-degree relatives, and by treatment received . The univariate and multivariate hazard ratios associated with each of these factors are presented in P=0.008).In the entire sample, the 5-year actuarial risk of contralateral breast cancer was 13.1% (95% CI 10.3–15.9%), the 10-year risk was 22.0% (95% CI 19.2–26.8%), and the 15-year risk was 33.8% (95% CI 28.6–39.0%). The annual risk was 2.1%. The 5-, 10-, and 15-year cumulative risks of contralateral breast cancer were estimated for each age group and by mutation status . The risThe risk of contralateral breast cancer also depended on the family history of the patient. The effect of family history was only present for women whose initial breast cancer was diagnosed at age 49 years or below. For these women, the 15-year risks of contralateral breast cancer were estimated to be 33.4, 39.1, and 49.7% for women with zero, one, and two or more first-degree relatives with breast cancer diagnosed under age 50 years.P=0.002) (P=0.0006), but not for those with age 50 years or older  (P=0.01) (P=0.11)  underwent bilateral oophorectomy; 40 women before breast cancer diagnosis, 354 women within the year following breast cancer surgery, and 75 women at a later date (20 dates missing). Subjects with an oophorectomy had a significantly lower risk of contralateral breast cancer, compared with women without an oophorectomy  . The ris P=0.84) . For wom(P=0.01) . Oophore(P=0.11) . Among wP=0.03); but not after adjustment for age of diagnosis and the other treatments  compared with 19% risk for women over the age of 50 years at time of diagnosis . These findings are consistent with those of P=0.9) in the entire study population or in any subgroup. This finding contrasts with that of P=0.04). In our study, we also included deceased cases to eliminate the potential for survival bias.In the current study, we did not observe a protective effect against contralateral breast cancer associated with chemotherapy use  and was significant for those with a BRCA1 mutation (53% reduction in risk). A recent meta-analysis was published of eight studies that estimated the risk of first primary breast cancer in BRCA1/2 mutation carriers who were treated with prophylactic oophorectomy relative to carriers who had intact ovaries  would be non-carriers. This suggests that in 810 breast cancer cases included in this study, four of the cases (0.4%) may not have had a BRCA mutation. There are limitations as well. Ideally, we would conduct a prospective study based on unselected cases.In summary, women with a BRCA mutation have a high risk of developing a contralateral breast cancer after the diagnosis of a first breast cancer. This risk is highest for women diagnosed with breast cancer under the age of 50 years and for those with multiple first-degree relatives affected with breast cancer. For women with intact ovaries, two or more first-degree relatives with breast cancer and who were diagnosed at age 49 years or below, the 15-year risk of contralateral breast cancer was 68%. For these high-risk women, both oophorectomy and contralateral mastectomy should be discussed as a component of the treatment plan to reduce the risk of second primary breast cancer and to prevent ovarian cancer.


Breast density, measured as the proportion of the breast occupied by fibroglandular tissue in a mammogram, is a strong and modifiable risk factor for breast cancer. Area-based estimates made by expert observers are a practical approach, but are subjective. Here we investigate repeatability of visual assessment of percentage breast density.Seven mammographic film readers re-assessed the density of 100 normal full-field digital mammogram cases for which they had made density estimates at least 1 year previously as part of the Predicting Risk of Cancer at Screening (PROCAS) study. The mammograms for a given reader were selected to show a range in density, by randomly sampling 10 cases from each decile of density assessed by that reader. They were reviewed in similar reading conditions on both occasions using a visual analogue scale to record the assessments.For the majority of readers the difference in mean density between the two sets of readings was less than 6%, but the largest discrepancy between means was 14.7%. Bland-Altman plots were generated for each reader and showed considerable variation between readings on the two occasions. At best, the limits of agreement were -12.46% to +17.02%, and at worst they were -14.50% to +40.98%. The largest difference between first and second readings for each reader ranged from 26 to 65%.Although density estimates made by a subset of these readers have been strongly related to cancer risk, the variability in reproducibility calls into question the usefulness of subjective assessment without prior evaluation of reader performance.


Herein we present the results of two related investigations. The first study determined if concentrations in breast nipple discharge (ND) of two proteins  predicted the presence of breast atypia and cancer in pre- and/or postmenopausal women requiring surgery because of a suspicious breast lesion. The second study assessed if these proteins increased the predictive ability of a carbohydrate  which we previously demonstrated predicted the presence of disease in postmenopausal women requiring surgery.In the first study we prospectively enrolled 79 participants from whom we collected ND, measured uPA and PAI-1 and correlated expression with pathologic findings. In the second study we analyzed 35  ND samples collected from different participants requiring breast surgery, all of whom also had TF results.uPA expression was higher in pre- and PAI-1 in postmenopausal women with 1) cancer (DCIS or invasive) vs. either no cancer , or benign pathology ; and 2) abnormal (atypia or cancer) versus benign pathology . High uPA and PAI-1 concentrations and age were independent predictors of disease in premenopausal women, with an area under the curve (AUC) of 83-87% when comparing diseased vs. benign pathology. uPA, TF, and age correctly classified 35 pre- and postmenopausal women as having disease or not 84-91% of the time, whereas combining uPA+PAI-1+TF correctly classified 24 women 97-100% of the time.uPA and PAI-1 concentrations in ND were higher in women with atypia and cancer compared to women with benign disease. Combining uPA, PAI-1 and TF in the assessment of women requiring diagnostic breast surgery maximized disease prediction. The assessment of these markers may prove useful in early breast cancer detection. Cancer cell invasion and metastasis requires the degradation of the extracellular matrix (ECM) and basement membrane. This process is accomplished by several proteins, including those of the plasminogen activator (PA) system. Urokinase-type PA (uPA) plays a key role in ECM degradation. In women with breast cancer, uPA appears to promote cancer invasion and metastasis  through Plasminogen activator activity is inhibited by plasminogen activator inhibitor-1 (PAI-1) . PAI-1 pIn a pooled analysis of 8377 breast cancer patients, higher uPA and PAI-1 levels in tumor tissue were related to worse prognosis . On the In this study we demonstrate in a specific population of participants (women with a suspicious breast lesion which required biopsy to exclude cancer) that the expression of uPA and PAI-1 is altered in women with breast atypia and cancer. Since these markers were more predictive of disease in pre- than in post-menopausal women, and were not perfectly accurate in predicting the presence of disease in either menopausal group, we wondered if additional marker(s), ideally one or more that was especially predictive of postmenopausal disease, would improve the ability of uPA and PAI-1 to predict the presence of breast atypia or cancer.We previously reported  that thein situ (DCIS) or invasive cancer; "no cancer" as histopathology which was normal, usual hyperplasia, and/or atypical hyperplasia; "benign" as histopathology containing normal findings and/or usual hyperplasia; and "abnormal" as histopatology demonstrating atypical hyperplasia, DCIS, and/or invasive cancer.After receiving Institutional Review Board approval, we obtained informed consent, prospectively enrolled participants and collected 79 ND samples in Grand Forks, North Dakota, at the University of North Dakota, in Columbia, MO, at the Ellis Fischel Cancer Center and in London, UK, at the Royal Marsden Cancer Center. Participants were enrolled from January 2008 to February 2010. Details of TF sample collection and analysis were previously reported [ND (1-10 μL) was obtained from the breast with a lesion prior to surgery. Lesions included women with 1) pathologic nipple discharge (PND); 2) a suspicious lesion identified on imaging, be it mammogram, ultrasound or breast magnetic resonance imaging; and/or 3) a palpable lesion that was not a simple cyst. Samples were collected as described previously . BrieflyThe portion of the capillary tube containing the sample was introduced into a 1.7 mL Eppendorf tube and 100 μL of a 0.1 mol/L solution of sodium bicarbonate (pH 7.8) added. The capillary tube was then crushed with a glass rod and the mixture vortexed to disperse the sample. The crushed capillary tube was left in the bicarbonate buffer overnight at 4°C, and the mixture then centrifuged  and the supernatant used without further dilution.uPA and PAI-1. ELISA kits for uPA and PAI-1 were obtained from American Diagnostica, Inc. . Levels of these two markers in ND samples were determined according to the manufacturer's instructions. Briefly, 100 uL of standards, samples and blanks were pipetted into microplate wells coated with monoclonal antibodies respectively specific for uPA and PAI-1 and incubated overnight at 4°C. After washing (× 4), enzyme-linked antibodies specific for each analyte were added to the wells and incubated for 1 h at room temperature. The wells were washed again, diluted enzyme conjugate (streptavidin conjugated horseradish peroxidase) was pipetted into the wells, incubated , then washed again. Substrate reagent was added to each well, followed by a stop solution (0.5 M sulfuric acid). Absorbance was measured with a microplate reader. Detection limits were 10 pg/mL for uPA and 50 pg/mL for PAI-1.TF. The evaluation of this carbohydrate was previously reported [reported .Biomarker levels for uPA and PAI-1 in the ND samples were heavily skewed and not normally distributed. We therefore described and analyzed biomarker levels using medians and log-transformed (Log10) means to achieve normal distributions. We performed a Shapiro-Wilk test to determine the goodness-of-fit of the data to a normal distribution. None of the log-transformed variables were significantly different from a normal distribution. Unpaired t-tests were calculated to determine significant differences between log10 means of the biomarkers. Logistic regression analyses are based on log10 transformed data. Three logistic regression models were used to examine the effects of uPA and PAI-1 in predicting (a) cancer vs. no cancer diagnosis; (b) cancer vs. benign diagnosis; and (c) abnormal vs. benign diagnosis. Various demographic factors, include age, menopausal status, and hormone replacement, were included in the logistic models as covariates. Because menopausal status was a near significant predictor in all three overall logistic models, follow-up logistic models were conducted separately for pre- and for post-menopausal women, controlling for age. Receiver operating characteristic (ROC) curves were calculated based on the logistic regression results for the premenopausal women's data, and area under the curve (AUC) values are presented. The optimal subset of all variants, uPA, PAI-1, TF, Tn, and covariates was selected to achieve the best AUC value.NAF was successfully collected in 90% (76/84) and PND in 100% (3/3) women. Among the 79 women evaluated for uPA and PAI-1 expression, 41 (51.9%) were postmenopausal. Age ranged from 21 to 82 years, with a median age in pre- and postmenopausal women of 44 and 61, respectively. Among the 35 women evaluated for TF, uPA +/- PAI-1 expression, 17 (48.6%) were postmenopausal. Age ranged from 26 to 77 years, with a median age in pre- and postmenopausal women of 43.5 and 61 years, respectively.ND volume ranged from 4 to 521 μl, with a mean of 60.9 and median of 29 μl. ND volume did not vary by pathologic diagnosis nor by fluid type (NAF vs. PND). We first determined if uPA and PAI-1 expression in women requiring diagnostic biopsy due to PND differed from expression in women requiring biopsy who did not have PND. Results of logistic regression analyses conducted using samples from women with PND and NAF showed that fluid type did not affect expression levels, with p > 0.5 for all biomarkers. Therefore, all reported analyses include both PND and NAF samples (N = 79).uPA concentration was higher Table  in womenUnivariate analysis indicated that uPA concentration was more predictive of disease in premenopausal women, and PAI-1 in postmenopausal women . We then measured uPA and PAI-1 in NAF collected contemporaneously with that used to measure TF. For all of these analyses, age was in the predictive model. TF+uPA predicted disease in both pre- and postmenopausal women with 84-92% accuracy. When TF, uPA, and PAI-1 were combined, the predictive ability approached 100%  which together appear highly predictive of the presence of cancer or precancer in women requiring breast biopsy to exclude disease. The markers appear to be highly predictive of disease, regardless of whether or not PND was present. The NAF markers appear complementary, as two of the biomarkers (TF and PAI-1) are more predictive in post- than premenopausal women, whereas uPA is more predictive in premenopausal women.The study had limitations. The first limitation of the study was the small sample size for the analysis of all three markers. Second, TF was analyzed at a different time than uPA and PAI-1. How much the age of the samples influenced marker expression is uncertain, though we did determine that uPA and PAI-1 expression was similar between the groups of 79 and 24 samples. A validation study in a larger sample size would help to determine the predictive ability of uPA, PAI-1 and TF would help to address these limitations.In women requiring breast biopsy to exclude disease, combining ND biomarkers that were each predictive of the presence of atypia or cancer improved the prediction of both atypia and cancer, regardless of whether or not PND was present. A validation study is warranted to determine the clinical use of these noninvasive markers.in situ; H: Usual hyperplasia; ICC: Intraclass correlation coefficient; NAF: Nipple aspirate fluid; ND: Nipple discharge; OR: Odds ratio; PND: Pathologic nipple discharge; ROC: Receiver operating characteristic.Abs: Absorbance; AH: Atypical hyperplasia; AUC: Area under the curve; CV: Coefficient of variation; DCIS: Ductal carcinoma Dr. Sauter is a consultant for Atossa Genetics, Inc.Dr. Qin performed specimen analysis. Drs. Gui and Twelves enrolled participants and critically reviewed the manuscript. Dr. Zhang performed the statistical analysis. Ms. Kliethermes performed data entry, managed the database and revised the manuscript. Dr. Sauter designed and oversaw all aspects of the study, enrolled participants, and prepared the preliminary and final drafts of the manuscript. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2407/12/52/prepub


We analyzed data from the Japan Collaborative Cohort Study (36 164 women aged 40–79 years at baseline in 1988–1990 with no previous diagnosis of breast cancer and available information on weight and height) to examine the association between baseline body mass index (BMI)/weight gain from age 20 years and breast cancer risk in a non-Western population.The participants were followed prospectively from enrollment until 1999–2003 (median follow-up: 12.3 years). During follow-up, breast cancer incidence was mainly confirmed through record linkage to population-based cancer registries. A Cox proportional hazards model was used to calculate hazard ratios (HRs) and 95% CIs for the association between breast cancer risk and body size.P < 0.0001). Risk was significantly increased among women with a BMI of 24 or higher  as compared with women with a BMI of 20 to 23.9. Weight gain after age 20 years and consequent overweight/obesity were combined risk factors for postmenopausal breast cancer risk. This combined effect was stronger among women aged 60 years or older. However, the HRs were not significant in premenopausal women.In 397 644.1 person-years of follow-up, we identified 234 breast cancer cases. Among postmenopausal women, the adjusted HR increased with BMI, with a significant linear trend (Our findings support the hypothesis that weight gain and consequent overweight/obesity are combined risk factors for breast cancer among postmenopausal women, particularly those aged 60 years or older. The recent continuous increase in breast cancer incidence has been an important public health concern in Japan, and the attention devoted to obesity/weight gain as a risk factor for breast cancer has also increased. Obesity is a well-known risk factor for postmenopausal breast cancer.2–4 Numerous epidemiologic studies have reported positive associations between obesity and breast cancer risk among white,5–10 African-American,11–13 and East Asian women.14–17 Furthermore, weight gain has been reported as an independent risk factor.8,9,11,17–21 Several studies have reported an inverse association between body weight in early adulthood and breast cancer incidence.17,19,20 However, the association has been somewhat inconsistent among premenopausal women. Obesity is associated with a decreased risk of breast cancer among white women,4,10,22–24 although accumulating evidence suggests that the inverse association is limited to women with estrogen receptor- and progesterone receptor-positive tumors.25–28 Studies of non-white racial/ethnic groups are more limited, and the results are mixed.Since the early 1990s, breast cancer has been the most frequently diagnosed cancer in Japanese women.To assist in cancer prevention, we analyzed data from a large cohort study—the Japan Collaborative Cohort (JACC) Study—which included 64 327 Japanese women, to examine the association of baseline body mass index (BMI)/weight gain with breast cancer risk, considering menopausal status at baseline. We also investigated the interaction of age on this association.29,30 In brief, the JACC Study was initiated in 1988–1990 and included 110 792 individuals (46 465 men and 64 327 women) aged 40 to 79 years from 45 areas throughout Japan. All participants were subsequently followed for all-cause mortality. In addition, study participants living in 24 areas with cancer registry systems were followed for cancer incidence.We analyzed data from the JACC Study, a prospective cohort study that evaluated cancer risk associated with lifestyle factors among the Japanese population. The study has been described in detail previously.Of the 64 327 women in the baseline cohort, 38 720 lived in the 24 areas where data on cancer incidence were available. The present study excluded 248 women who reported a previous diagnosis of breast cancer and 2308 women who did not provide information on height or weight at baseline. Thus, 36 164 women were included in the present analysis.Informed consent was obtained from the participants in the form of signatures on the cover pages of the questionnaires, with the exception of those in a few study areas where informed consent was provided at the group level after the aims and data confidentiality had been explained to community leaders. The Ethics Board of Sapporo Medical University approved our study.2). Information regarding weight and height was obtained from the self-reported questionnaire. Change in weight from age 20 years to the baseline measurement was calculated as the difference in the reported values at baseline among 20 418 women whose information on weight at age 20 years was available. We did not use BMI for age 20 years because we did not have access to height information at that age.As a relative indicator of body weight, BMI was calculated as weight in kilograms divided by the square of the height in meters  participants were lost to follow-up due to moving out of their designated study areas. Among the 234 breast cancer cases, no information on diagnosis was available for 13 (5.6%), ie, they were identified with death certification only (DCO). The world standard for DCO in cancer registration is less than 10%. The mortality-to-incidence ratio for breast cancer was 0.262 (58/221) in the cohort covered by cancer registries, which was within the range calculated using available data from population-based cancer registries in Japan (0.20–0.30). We estimated that 36.5 cases of incident breast cancer were not included in the cancer registries.31: less than 18.5, 18.5–19.9, 20–23.9, 24–28.9, and 29 kg/m2 or higher. Furthermore, BMI was entered directly to evaluate the linear trend of relative weight. The effect of age on the association between BMI and breast cancer risk was examined by analyzing the relationship between age and BMI. Finally, to investigate the combined effect of baseline BMI and weight change from age 20 years, we recategorized the participants into 4 groups using the following cutoff points: baseline BMI less than 24 kg/m2 and weight gain of less than 10 kg from age 20 years to the baseline measurement.For each cohort subject, person-years of follow-up were counted as time from enrollment to diagnosis of breast cancer, death from any cause, or end of follow-up (1999–2003), whichever occurred first. For breast cancer cases ascertained only by death certificates, person-years of follow-up were calculated from enrollment to death from breast cancer. Those who died from causes other than breast cancer or who moved out of the study areas were treated as censored cases. We used a Cox proportional hazards model to estimate hazard ratios (HRs) and 95% CIs for the association of breast cancer risk with baseline BMI/weight change. Women were divided into 5 categories, using baseline BMI 32 . All analyses were performed with regard to menopausal status and stratified by 6 study areas .We evaluated the association using age-adjusted and multivariable models with adjustment for age (using 10-year age groups), tobacco smoking , alcohol consumption , age at menarche , education level , parity , age at first birth , menopausal status , use of exogenous female hormone , first-degree family history of breast cancer , and physical activity categoriesP values were 2-sided, and a P value less than 0.05 was considered to indicate statistical significance. All regression analyses were performed using the PROC PHREG procedure of SAS Version 9.1 . Study areas were not incorporated in the Cox model with other potential confounders but were adjusted for using the strata option in the PHREG procedure.We repeated the analysis after excluding the first 2 years of follow-up, during which 38 cases of breast cancer were diagnosed. All 2, respectively. In 397 644.1 person-years of follow-up , we identified 234 breast cancer cases. Table Average age and BMI (SD) at baseline of the 36 164 women were 57.8 (10.0) years and 22.9 (3.1) kg/mP < 0.0001). Furthermore, significantly increased risk was observed among women with a BMI of 24 or higher  as compared with those with a BMI of 20 to 23.9. The adjusted HRs per 5-kg/m2 increment in BMI among pre- and postmenopausal women were 0.95 (95% CI: 0.60–1.50) and 1.68 (95% CI: 1.34–2.01), respectively.Table 2 increment in BMI in younger (40–59 years) and older (60–79 years) age groups. The older group had a higher HR  than the younger group  for a 5-kg/m2 increment of BMI, after adjustment for potential confounders.To observe the effect of age on the association between BMI and breast cancer risk among postmenopausal women, we calculated the HR for a 5-kg/mAn effect of weight gain between age 20 years and baseline on breast cancer risk was observed only among postmenopausal women. The HR (95% CI) for 1 increment of weight gain was 1.04 (1.01–1.07). Among premenopausal women it was 0.99 (0.94–1.04) and not significant.The combinatorial effect of baseline BMI and weight change between age 20 years and baseline was examined to evaluate the effect of these factors separately Table . In premTo our knowledge, this is the first prospective report from Japan on the association between obesity/weight gain and breast cancer risk by age group. Our findings revealed a significant association between BMI/weight gain and postmenopausal breast cancer risk, particularly among older women. For postmenopausal women, especially those aged 60 years or older, weight gain after age 20 years and consequent overweight/obesity were identified as combined risk factors for breast cancer, after adjusting for potential confounders. In other words, being overweight or obese at baseline was a much greater risk factor among women who were postmenopausal, were aged 60 years or older, and had gained at least 10 kg from age 20 years to baseline.2 increment in BMI in the present study (1.68) was slightly higher than the summary risk ratios from a meta-analysis4 of studies conducted in the Asia-Pacific (1.31), North America (1.15), and Europe and Australia (1.09). Breast cancer prevention via weight control is expected to be more effective among postmenopausal women in the Asia-Pacific region. With regard to cancer pathogenesis, the increased risk in overweight/obese postmenopausal women is due to the fact that adipose tissue is the major source of estrogenic hormones after menopause.33,34 Furthermore, our results conform with those of an earlier report showing that adult weight gain might be better than cross-sectional BMI as an adiposity index.35Our results for postmenopausal women are consistent with those obtained in a number of studies worldwide. The adjusted HR per 5-kg/m36,37 Another hypothesis is that there is greater clearance of estrogen by the liver in young overweight women.38 These hypotheses are strengthened by results from studies suggesting that the inverse associations are limited to women with tumors that are estrogen receptor- and progesterone receptor-positive.25–28 Thus, the heterogeneity of pathologic types among premenopausal breast cancer weakens the association and possibly explains the inconsistent results among non-white racial/ethnic groups. This heterogeneity of cancer etiology in relation to BMI and receptor type makes cancer prevention in premenopausal women difficult and of less practical importance. Further investigations of cancer pathogenesis are needed among non-white racial/ethnic groups.In contrast, we did not observe any significant association between BMI/weight change and breast cancer risk among premenopausal women. In our cohort, age at baseline was 40 years or older; thus, follow-up did not completely cover the premenopausal period. A previous study reported an inverse association between BMI and breast cancer risk among white women. One hypothesis is that young overweight women are more likely to have anovulatory cycles with less cumulative exposure to endogenous estrogen.A major advantage of the present study was its prospective design, which may avoid the possibility of recall bias inherent to case-control studies. Moreover, information on other breast cancer risk factors was included, and potential confounding factors were controlled in analyses of the association.This study has some limitations that should be considered when interpreting our results. First, because we did not have updated information on menopausal status, which would modify the association between BMI/weight change and breast cancer, the possibility of misclassification of menopausal status at breast cancer onset should be considered. Such misclassification would be problematic in premenopausal women, since recently menopausal women would be misclassified as premenopausal during the follow-up period. Such misclassification could partly explain the inconsistent results from several studies of the association between body size and breast cancer among premenopausal women. Studies of younger women with updated information on menopausal status should be initiated among premenopausal women. However, this limitation is a minor concern for postmenopausal women. Changes during follow-up, especially those related to lifestyle, might alter the results. However, many risk factors, such as marriage status, number of children, and family history of breast cancer, would be unlikely to change after age 40. To our knowledge, substantial changes in risk factors for breast cancer related to BMI have not been reported.Second, because we used simple questionnaires at baseline only, we have data at only 2 time points, ie, age 20 years and baseline. We did not have data on the time period of weight gain, which would provide useful information for recommendations. Lack of information on weight gain around menopause would also weaken the association among premenopausal women. Furthermore, weight at age 20 years is retrospective information and may be systematically biased among women at extremes of body size. However, these data were obtained before breast cancer diagnosis, and therefore any misclassification is not likely to be differential.The accuracy of cancer identification in the present study was not ideal. We estimated that 36.5 cases of incident breast cancer were not included in our follow-up, and this number is not inconsiderable. However, these cases would be independent of body size; thus, estimated HRs would tend toward the null.In summary, our findings support the hypothesis that a weight gain of 10 kg or more and consequent overweight/obesity (BMI ≥24) are combined risk factors for breast cancer among Japanese postmenopausal women, particularly those aged 60 years or older. Thus, to prevent breast cancer, weight gain after age 20 years should be avoided and weight control should be increasingly emphasized with increasing age. The association between body size and premenopausal breast cancer was not clear in the present study and varies across studies; thus, optimal weight for breast cancer prevention cannot be specified at this time.Abstract in Japanese.


This study was undertaken to examine whether there is an association between parity and age at first birth and risk of death from brain cancer.The study cohort consisted of 1,292,462 women who had a first and singleton childbirth between Jan. 1, 1978 and Dec. 31, 1987. We tracked each woman from the time of their first childbirth to December 31, 2009, and their vital status was ascertained by linking records with the computerized mortality database. Cox proportional hazard regression models were used to estimate the hazard ratios (HR) of death from brain cancer associated with parity and age at first birth.There were 316 brain cancer deaths during 34,980,246 person-years of follow-up. The mortality rate of brain cancer was 0.90 cases per 100,000 person-years. The adjusted HR was 1.35 (95% CI= 0.91-2.01) for women who gave birth between 21 and 25, 1.61 (95% CI=1.05-2.45) for women who gave birth after 25 years of age, respectively, when compared with women who gave birth less than 20 years. A trend of increasing risk of brain cancer was seen with increasing age at first birth. The adjusted HR were 0.73 (95% CI= 0.53-0.99) for women who had 2 children, and 0.60 (95% CI =0.43-0.83) for women with 3 or more births, respectively, when compared with women who had given birth to only 1 child. There was a significant decreasing trend in the HRs of brain cancer with increasing parity.This study provides evidence that reproductive factors (parity and early age at first birth) may confer a protective effect on the risk of death from brain cancer. In Taiwan, brain cancer is the 15th leading cause of cancer mortality for males and the 13th for females[Little is known regarding the etiology of brain cancer. Ionizing radiation is the most consistently accepted causal risk factors for brain cancer, and thiThe incidence rate of glioma is about 1.5 times greater in men than in women. The higThe role of parity and age at first birth in the etiology of brain cancer in women have received only limited attention in the literature, and the results have been inconsistent. Some studies reported inverse associations between increasing parity and the risk of brain cancer-14 and oTo date, there are limited data regarding the relationship between reproductive factors and brain cancer. The above-mentioned epidemiologic studies have been conducted on high-incidence populations in Europe and North America and study results have not been consistent. The present study was carried out because no studies have been conducted on low-incidence populations such as Asians. We studied a cohort of women who experienced a first and singleton childbirth between January 1, 1978 and December 31, 1987 to explore further the association between parity, age at first birth and the risk of brain cancer in Taiwan.Registration of births is required by law in Taiwan. It is the responsibility of the parents or the family to register infant births at a local household registration office within 15 days. The Birth Registration System, which is managed by the Department of Interior, released computerized data on live births since 1978. The registration form, which requests information on maternal age, education, parity, gestational age, date of delivery, infant gender, and birth weight, is completed by the physician attending the delivery. Because most deliveries in Taiwan take place in either a hospital or clinic and the The study cohort consisted of all women with a record of a first and singleton childbirth in the Birth Register between January 1, 1978 and December 31, 1987. There were 1,333,312 first and singleton births occurred in Taiwan between 1978–1987. Information on any subsequent births was also retrieved from the Birth Register. Of the 1,399,312 primiparous women, 106,850 subjects were excluded because data were missing on at least one variable such as maternal age , years of schooling (n=382), marital status (n=2665), or birth place (n=4535). These exclusion left 1,292,462 women with complete information for the analysis. Their details have already been described in an earlier publication.Each woman has her own unique personal identification number. Using this number, we tracked each woman from the time of their first childbirth to December 31, 2009, and their vital status was ascertained by linking records with the computerized mortality database, identifying the date of any deaths occurring in this cohort. Of the 1,292,462 women followed, none had a missing personal identification number; therefore, all cohort members were followed up. Since it is mandatory to register death certificates at local household registration offices, the mortality statistics in Taiwan were considered to be highly accurate and complete.The person-years of follow-up for each woman was calculated from the date of first childbirth to the date of death or December 31, 2009. Death rates were calculated by dividing the number of deaths from brain cancer by the number of person-years of follow-up. Cox proportional hazard regression models were used to estimate the hazard ratios (HRs) of death from brain cancer associated with parity (the number of children recorded in the last childbirth record of each woman registered during follow-up) and age at first childbirth. The 95% confidence intervals (CIs) for the RRs were also calculated. Brain cancer is defined according to the International Classification of Disease, Injury, and Causes of Death (9th revision) (ICD code 191). It includes malignant neoplasm of the cerebrum, except lobes and ventricles (191.0); malignant neoplasm of the frontal lobe (191.1); malignant neoplasm of the temporal lobe (191.2); malignant neoplasm of the parietal lobe (191.3); malignant neoplasm of the occipital lobe (191.4); malignant neoplasm of the ventricle (191.5); malignant neoplasm of the cerebellum (191.6); malignant neoplasm of the brain stem (191.7); malignant neoplasm of other parts of brain (191.8); and malignant neoplasm of brain, unspecified (191.9). The variables in the final model included age at first childbirth , parity , marital status , years of schooling , and birth place . The proportional hazards assumption was assessed for all above-mentioned variables, and no violations were observed. Analyses were performed using the SAS statistical package . All statistical tests were two-sided; p values of less than < 0.05 were considered to be statistically significant.Altogether 1,292,462 primiparous women with complete information were included in the analysis. A total of 34,980,246 person-years were observed during the follow-up period from the time of their first childbirth to December 31, 2009. There were 316 brain cancer deaths, yielding a mortality rate of 0.90 cases per 100,000 person-years.Table The multivariate-adjusted HR and 95% CIs are shown in Table After adjustment for age at first birth, marital status, years of schooling, and birth place, the adjusted HR were 0.73 (95% CI= 0.53-0.99) for women who had 2 children, and 0.60 (95% CI =0.43-0.83) for women with 3 or more births, respectively, when compared with women who had given birth to only 1 child. There was a statistically significant decreasing trend in the adjusted HRs for brain cancer with increasing parity (p for trend=0.003).To our knowledge, this is the largest cohort n=1,292,462 women) published to date to examine the relationship between reproductive factors (parity and age at first birth) and brain cancer risk. In this prospective cohort study, we found a statistically significant decreased risk of brain cancer with increasing parity. Our findings of a reduced risk of brain cancer associated with higher parity is in agreement with previous studies-14, but 62 women The mechanism by which increased parity may confer protection against the future development of brain cancer in women remains unclear. Pregnancy elevates serum estrogen levels approximately 100-fold. IncreasIn the present prospective cohort study, we found that risk of brain cancer increased with increasing age at first birth after adjusting for parity. Our finding of a positive association between age at first birth and brain cancer risk agrees with one previous study but not To our knowledge, ours is the first prospective cohort study to report a protective effect of parity and early age at first birth on the subsequent risk of death from brain cancer. However, these results are inconsistent with most studies published to date. The reasons for these differences in findings are unknown. Only three prospective studies have examined the relationship between parity and age at first birth and brain cancer risk,19,22. MMortality data have been widely used to generate epidemiologic hypotheses, despite their inherent limitations. The completeness and accuracy of the death registration system should be evaluated before any conclusion based on the mortality analysis is made. In the event of a death in Taiwan, the decedent’s family is required to obtain a death certificate from the hospital or local community clinic, which then must be submitted to the household registration office in order to cancel the decedent’s household registration. The death certificate is required in order to have the decedent’s body buried or cremated. Death certificates must be completed by physicians in Taiwan. It is also mandatory to register all deaths at local household registration offices, the death registration is accurate, reliable and complete.Taiwan is a small island with a convenient communication network. It is believed that all brain cancer cases had access to medical care. Mortality data rather than data on inpatient cases was used to assess the association between parity, age at first birth, and brain cancer in this study. The mortality of a disease is a function of its incidence and fatality. The 5-year survival rate for brain cancer has been reported to be as low as 30% in United States. Deaths Several studies have found that the use of OCs was associated with a reduced risk of brain cancer,17,21. OSmoking was associated with the incidence of brain tumors-42. TherIonizing radiation, occupational exposure to solvents, and electromagnetic fields are the most important risk factors for brain cancer. There iSome potential limitations of this study needed to be noted. First, our computerized death-certificate database does not specify the tumor type and the histological classification for the brain cancer cases. We therefore could not estimate the risk of brain cancer in more detail in relation to risk for specific brain cancer subtypes. Second, data on the accuracy of brain cancer diagnosis are not available in Taiwan and misclassification is possible. However, this misclassification is likely to be nondifferential  and therefore would tend to underestimate rather overestimate the true association. Third, Taiwan’s vital records and birth registration system covers only live births and exclude stillbirths and abortions. We were therefore unable to examine the possible role of gravidity on the brain cancer risk. Fourth, by design, our study focused solely on mortality among parous women. We were unable to examine the possible role of nulliparity on the risk of brain cancer. The generalizability of our findings is thus limited. Fifth, although the number of person-years are large, the number of of brain cancer deaths is relatively small (n=316). The power of this study was therefore relatively limited. The possibility that this is a chance finding also needs to be considered.We found that there was a trend for increasing parity to be associated with decreasing risk for brain cancer. In addition, we found that risk for brain cancer increased with increasing age at first birth. This study provides evidence that reproductive factors (parity and early age at first birth) may confer a protective effect on the risk of brain cancer.The authors declare that they have no competing interests.HFC wrote the manuscript. CCC and SST provided essential insight into the interpretation of the results. SCH did the statistical analysis. CYY contributed to study design and interpretation of the data. He had full access to all of the data in the study and took responsibility for the integrity of the data and the accuracy of the data analysis. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2458/12/857/prepub