This case-control study assesses the association of a genetic risk score based on variants in a DNA repair pathway with risk of a second primary contralateral breast cancer after radiation therapy for a first breast cancer. Is a genetic risk score comprising variants in a DNA repair pathway associated with risk of developing a second primary contralateral breast cancer among women who underwent radiation therapy for a primary breast cancer?In this case-control study including 3732 women who received a diagnosis for a first invasive local or regional breast cancer when they were younger than 55 years, a genetic risk score comprising variants in a DNA repair pathway was associated with increased risk of a subsequent radiation-associated contralateral breast cancer. Among younger women with a high genetic risk score, the attributable increased risk for contralateral breast cancer associated with stray radiation dose was 28%.This genetic risk score may be a helpful tool to guide treatment for young women with breast cancer. Radiation therapy for breast cancer is associated with increased risk of a second primary contralateral breast cancer, but the genetic factors modifying this association are not well understood.To determine whether a genetic risk score comprising single nucleotide polymorphisms in the nonhomologous end-joining DNA repair pathway is associated with radiation-associated contralateral breast cancer.This case-control study included a case group of women with contralateral breast cancer that was diagnosed at least 1 year after a first primary breast cancer who were individually matched to a control group of women with unilateral breast cancer. Inclusion criteria were receiving a first invasive breast cancer diagnosis prior to age 55 years between 1985 and 2008. Women were recruited through 8 population-based cancer registries in the United States, Canada, and Denmark as part of the Women’s Environment, Cancer, and Radiation Epidemiology Studies I (November 2000 to August 2004) and II (March 2010 to December 2012). Data analysis was conducted from July 2017 to August 2019.Stray radiation dose to the contralateral breast during radiation therapy for the first breast cancer. A novel genetic risk score comprised of genetic variants in the nonhomologous end-joining DNA repair pathway was considered the potential effect modifier, dichotomized as high risk if the score was above the median of 74 and low risk if the score was at or below the median.The main outcome was risk of contralateral breast cancer associated with stray radiation dose stratified by genetic risk score, age, and latency.A total of 5953 women were approached for study participation, and 3732 women (62.7%) agreed to participate. The median (range) age at first diagnosis was 46 (23-54) years. After 5 years of latency or more, among women who received the first diagnosis when they were younger than 40 years, exposure to 1.0 Gy  or more of stray radiation was associated with a 2-fold increased risk of contralateral breast cancer compared with women who were not exposed . The risk was higher among women with a genetic risk score above the median , and there was no association among women with a genetic risk score below the median . Among younger women with a high genetic risk score, the attributable increased risk for contralateral breast cancer associated with stray radiation dose was 28%.This study found an increased risk of contralateral breast cancer that was attributable to stray radiation exposure among women with a high genetic risk score and who received a first breast cancer diagnosis when they were younger than 40 years after 5 years or more of latency. This genetic risk score may help guide treatment and surveillance for women with breast cancer. Radiation exposure induces a variety of lesions in DNA, the most dangerous of which are DNA double strand breaks. In humans, double strand breaks are repaired primarily by the relatively error-prone nonhomologous end-joining (NHEJ) pathway, whereas the less error-prone homologous recombination pathway is active mainly during the G2 phase of actively cycling cells.15 Research has shown that mutations in DNA damage response genes are associated with hypersensitivity to ionizing radiation and a high incidence of cancer.16 Additionally, a 2007 study17 among Taiwanese women found that variants in 2 genes within the NHEJ pathway were associated with breast cancer risk. As breast cancer is the most common malignant neoplasm among women in the United States18 and radiation therapy is used to treat more than half of women with breast cancer,19 identifying risk factors for radiation-associated contralateral breast cancer is an important issue.Survivors of invasive breast cancer have a high risk of developing asynchronous contralateral breast cancer.22 A recent genome-wide association study of survivors of childhood cancers treated with radiation therapy identified several loci that may interact with radiation exposure to increase risk of subsequent breast cancer,23 but it is uncertain whether risk of cancer after radiation therapy among children can be generalized to risk of cancer after radiation therapy among adults. Genetic risk scores (GRSs), which aggregate the associations of many genetic variants, are a promising method to identify associations that are not detectable for individual variants.24 Therefore, we genotyped single nucleotide polymorphisms (SNPs) in women with contralateral breast cancer (case group) and individually matched women with unilateral breast cancer (control group) in the Women’s Environmental Cancer and Radiation Epidemiology (WECARE) Study to develop a GRS that captures variation in risk of contralateral breast cancer associated with stray radiation exposure. We considered several approaches to develop this GRS, including aggregating SNPs associated with contralateral breast cancer in the WECARE Study, aggregating SNPs associated with contralateral breast cancer in the literature,24 and aggregating SNPs located in or near genes in specific DNA damage response pathways.To date, only rare, high-penetrance genetic mutations have been evaluated in radiation-associated contralateral breast cancer, to our knowledge.20; we hypothesized that having a greater number of alleles associated with cancer risk in the NHEJ pathway would be associated with a greater risk of contralateral breast cancer associated with exposure to stray radiation. By aggregating SNPs in and adjacent to the 7 genes encoding the NHEJ pathway, we assessed whether a high score on this pathway-specific NHEJ GRS was associated with a high risk of contralateral breast cancer subsequent to stray radiation exposure in the WECARE Study.In this study, we focused on a GRS comprising genes in the NHEJ DNA damage response pathway, which is the most common pathway for DNA double strand break response in humans in all phases of the cell cycle. The rationale for focusing on the NHEJ pathway is that genetic variation in genes involved in DNA damage response is associated with cancer risk25 which recruited from November 2000 to August 2004, and the WECARE Study II,26 which recruited from March 2010 to December 2012. Participants included women who had a first invasive local or regional breast cancer that was diagnosed between 1985 and 2008 and a subsequent primary invasive cancer or carcinoma in situ in the contralateral breast  with no cancers in the intervening period. Breast cancer primary status was further confirmed by the Surveillance, Epidemiology, and End Results program for participants in the United States and by medical record review after identification in the Danish Cancer Registry for women in Denmark. The control group included women who had received a first primary local or regional breast cancer and no subsequent contralateral breast cancer. Women in the control group were individually matched with women in the case group at a ratio of 2 to 1 in the WECARE Study I25 and a ratio of 1 to 1 in the WECARE Study II26 by age when they received the first breast cancer diagnosis (5-year strata), diagnosis year (4-year strata), cancer registry region, and self-reported race/ethnicity. The at-risk period for a woman with contralateral breast cancer was the time between when she received the first breast cancer diagnosis to when she received the second breast cancer diagnoses. Women in the control group were randomly selected from women in the risk set who were living and had not been diagnosed with any cancer during the interval after they received the first diagnosis corresponding to the length of the at-risk period of the matching woman with contralateral breast cancer. In the WECARE Study I,25 case-control sets were counter-matched such that exactly 2 members of the triad had cancer registry–reported radiation treatment for the primary breast cancer, which increased power to detect gene-environment interactions for contralateral breast cancer. A summary of enrollment criteria for each phase of the WECARE Study is provided in eTable 1 in the The WECARE Study is a multicenter, population-based, case-control study of women with contralateral breast cancer as the case group and individually-matched women with unilateral breast cancer as the control group. Participants were recruited through 8 population-based cancer registries in the United States, Canada, and Denmark in 2 phases: the WECARE Study I,25 and saliva in WECARE Study II26). Data on tumor characteristics and treatment were obtained from cancer registries and by abstraction of medical records. We followed the Strengthening the Reporting of Observational Studies in Epidemiology (STROBE) reporting guideline.The study was approved at each participating research center by its institutional review board and, in Denmark, additionally by the ethics committee system. Participants gave verbal informed consent for a telephone interview that collected sociodemographic and clinical data, including a detailed breast cancer risk-factor history. After obtaining written informed consent, biological specimens were collected for DNA extraction and genotyping  and the areola region. The dose estimates were based on individual radiation parameters  from available information, including radiation therapy records, summary notes, and abstracted physician correspondence from medical records. Stray doses to the contralateral breast quadrants and areola were measured using lithium fluoride powder thermoluminescent dosimeters placed in tissue-equivalent phantoms, molded on women in treatment position. All reported doses are location-specific to estimate the radiation dose received by the affected quadrants or areola of the woman’s contralateral breast during treatment for the first breast cancer and the corresponding region in the unaffected breast of her matched control, as previously described.14Radiation dose estimation methods have been detailed previously.25 SNPs were genotyped using the HumanOmni1-Quad BeadChip platform (Illumina). In the WECARE Study II,26 SNP genotyping was conducted using 2 custom oligonucleotide probe panels using the Infinium iSelect HD Custom BeadChip (Illumina). Quality control methods have been detailed previously.24 Self-reported race/ethnicity was confirmed by principal components analysis. To reduce the probability of confounding by population stratification, final analyses using genetic data were restricted to non-Hispanic white women.In the WECARE Study I,DCLRE1C (OMIM 602450), LIG4 (OMIM 606593), NHEJ1 (OMIM 611291), PRKDC (OMIM 600899), XRCC4 (OMIM 194363), XRCC5 (OMIM 194364), and XRCC6 (OMIM 152690). Ninety-three SNPs in the pathway passed quality control, and 24 SNPs were excluded owing to strong linkage disequilibrium (r2 > 0.5) with others in the pathway. To our knowledge, there are no reported significant associations of SNPs in the NHEJ pathway with radiation-associated second cancers. Therefore, for the final 69 SNPs, we determined the alleles associated with contralateral breast cancer risk by the directionality of the main association with contralateral breast cancer risk among all women in the WECARE Study. We chose to use the direction of association from the entire study population rather than only among those exposed to radiation to ensure that the NHEJ GRS was independent of the interaction between the score and radiation exposure. The final NHEJ GRS is the sum of risk alleles  across all 69 NHEJ SNPs. Women in the case group and the control group were classified as high NHEJ GRS if their score was greater than the median score of the all participants, 74  alleles, and low NHEJ GRS if their score was less than or equal to the median score.The NHEJ GRS is comprised of SNPs in or near the 7 genes in the NHEJ pathway: 27 To account for the counter-matched WECARE Study I25 design, statistical models included a log-weight offset term. The WECARE Study II26 participants, who were matched in pairs without counter-matching on radiation therapy, were assigned an offset equal to 1. All statistical models were adjusted for known and suspected contralateral breast cancer risk factors.28 Several important covariates had missing values for women in the case and control groups. To address missing data, we used the missing indicator method for a matched case-control study, which improves statistical efficiency relative to a complete-case analysis and is preferred to an unmatched analysis of incomplete pairs, which is vulnerable to additional confounding.29Multivariable-adjusted rate ratios (RRs) and corresponding 95% CIs were estimated using conditional logistic regression. Women in the control group were sampled from the failure time risk sets of the women in the case group, so the RR estimates are equivalent to those obtained in a proportional hazards model of cohort data.31 For each model of radiation dose and contralateral breast cancer risk, a single stratified model was developed to obtain stratum-specific estimates while avoiding overfitting within subgroups. Statistical tests for trend were performed across the dose categories.First, we estimated the association of radiation exposure with contralateral breast cancer risk. Models were developed separately for radiation therapy ever vs never exposure and for location-specific dose received to the contralateral breast . Stratified analyses were defined a priori by age at first diagnosis (<40 years vs ≥40 years) and latency between first and second cancers (<5 years vs ≥5 years), based on previous WECARE Study results.32For the primary analysis, we examined the joint associations of the NHEJ GRS and radiation dose with contralateral breast cancer risk, fitting conditional logistic regression models with stratification as described and with additional adjustment for 3 eigenvectors obtained in principal components analysis. We estimated the population-attributable risk fraction of contralateral breast cancer associated with radiation exposure from these results.24 that captures common genetic variation in known breast cancer susceptibility loci. This GRS was developed by aggregating SNPs associated with first breast cancer risk and was also associated with contralateral breast cancer risk in the WECARE Study.We assessed the robustness of the NHEJ GRS associations by adjusting for a GRS by Robson et alAll analyses were performed in SAS/STAT software version 9.4 (SAS Institute). All statistical tests were 2-sided, and statistical significance was set at less than .05.25 or WECARE Study II,26 of whom 3732 (63%) gave informed consent and participated. Participant characteristics are presented in A total of 5953 women were approached for enrollment in the WECARE Study IAll models were stratified by age at first diagnosis (<40 years vs ≥40 years) and latency (<5 years vs ≥5 years). Ever having received radiation therapy was associated with increased contralateral breast cancer risk among women younger than 40 years when they received the first diagnosis after a latency of 5 years or more  . The risP = .03). No association was observed between location-specific stray radiation dose and contralateral breast cancer risk in the overall study population , or other age or latency subgroups . There was a statistically significant dose trend (ubgroups .−4) (eTable 2 in the P = .03). In contrast, for women with an NHEJ GRS of 74 alleles or fewer in the same age and latency group, there was no association between radiation dose and contralateral breast cancer risk . No associations were found for women who received their first breast cancer diagnosis when they were 40 years or older. Based on these results, after a latency of 5 years or longer among women who received their first breast cancer diagnosis when they were younger than 40 years with a high NHEJ GRS, the population attributable risk fraction of contralateral breast cancer attributable to stray radiation exposure to the contralateral breast was 28%. The corresponding population attributable risk fraction among women who received their first diagnosis when they were younger than 40 years after a latency of 5 years or more with a low NHEJ GRS was 18%.Contralateral breast cancer risk was not associated with the interaction between individual SNPs in the NHEJ GRS and radiation dose after Bonferroni correction for multiple comparisons  (eTable 3 in the 33 for significant expression quantitative trait loci (eQTLs) affecting the genes encoding NHEJ components. For 5 of the 7 genes in the NHEJ pathway , we identified significant eQTLs associated with either increased or decreased expression of the corresponding NHEJ gene in multiple tissues. Notably, within each gene, significant eQTLs were always associated with a single direction of association .To explore the mechanism whereby variation in the NHEJ GRS is associated with risk of contralateral breast cancer associated with radiation exposure, we searched the NHEJ GRS SNPs against the Genotype-Tissue Expression (GTEx) database31 which we confirmed in the present study. We also found that women who received the first diagnosis when they were younger than 40 years and were exposed to at least 1.0 Gy of stray radiation had an increased risk of developing contralateral breast cancer after 5 years, but only if they carried many NHEJ GRS risk alleles, whereas those with a low NHEJ GRS had no significantly increased risk of contralateral breast cancer associated with radiation dose. For women who were 40 years or older when they received the first diagnosis, the risk of radiation-associated contralateral breast cancer was low regardless of latency time or NHEJ GRS. We estimated that, after a latency of 5 years or more among women with a high genetic risk who were younger than 40 years when they received the first diagnosis, 28% of the increased risk of contralateral breast cancer was attributable to stray radiation exposure. For younger women with breast cancer, these findings could influence decision-making for locoregional therapy. For example, young women with a high NHEJ GRS may consider partial-breast radiation therapy (rather than whole-breast radiation therapy when appropriate), opt for radiation therapy techniques that reduce integral dose , or decide for non-radiation therapy–based locoregional management .36 These findings may be especially important in younger women with medially located breast cancers, where the scatter dose to the contralateral breast is likely to be higher.36 Therefore, focusing on SNPs in the NHEJ pathway may prove useful for individualizing breast cancer treatment, particularly for women treated with radiation therapy when they are younger than 40 years.This study found that a substantial proportion of the risk of radiation-associated contralateral breast cancer was attributable to common genetic variants in the NHEJ pathway among younger women. Our pathway-specific GRS approach, based on our a priori hypothesis that variation in a DNA damage response pathway would be associated with radiation-associated contralateral breast cancer, allowed a valid and statistically powered evaluation of genetic variation and radiation-associated contralateral breast cancer risk. Radiation-associated contralateral breast cancer risk is characterized by an inverse association with age at exposure, a latency of effect, and a proportional association with radiation dose,31 We observed no associations between radiation exposure and contralateral breast cancer risk in the overall study population or among women who received the first diagnosis after age 40 years, affirming the importance of age- and latency-stratified analysis to identify women at increased risk of radiation-associated contralateral breast cancer.31The choice of age and latency cutoffs were based on evidence from previous studies suggesting the associations between cancer and radiation exposure are strongest for women exposed when younger than 40 years and that associations with radiation exposure are greatest after at least 5 years.Even after accounting for known breast cancer susceptibility SNPs (eTable 3 in the LIG4, NHEJ1, and XRCC5 genes while uniformly decreased for the XRCC4 and XRCC6 genes. The consistent association of multiple NHEJ GRS risk alleles with eQTLs in a single direction suggests that the NHEJ GRS may be capturing the effect of SNP alleles on the transcription of 1 or more genes in the NHEJ pathway. This supports the hypothesis that the variation in this pathway may alter double-stranded DNA damage response, thereby increasing the risk of tumor development. However, the results from GTEx are drawn from multiple tissues that may not be appropriate proxies for breast tissue and the impact of genetic variation in the overall NHEJ pathway is likely to be complex.In our gene-by-gene analysis in the high-risk subgroup (eTable 4 in the This study has limitations. First, although a large population of participants had quantified location-specific radiation dose, accurate dosimetry was not possible for 600 women with incomplete location or clinical data or multifocal tumors. Second, the NHEJ GRS was developed based on the directionality of SNP associations among non-Hispanic white women in the WECARE Study, which may limit generalizability to other populations. However, this does not reduce the validity of the score for this population. Third, it is not currently possible to replicate our NHEJ GRS findings in another data set, as no other study of contralateral breast cancer is large enough with the necessary biospecimens, long-term follow-up, and dosimetry and questionnaire data, to our knowledge. Strengths of the study include the large study population that allowed for subgroup analyses, individual matching to reduce confounding, and individualized dosimetry. Future studies should assess whether these results are consistent for other radiation-sensitive second primary cancers.In conclusion, this study found that as much as 28% of the risk of contralateral breast cancer was attributable to stray radiation exposure among women who received their first breast cancer diagnosis when they were younger than 40 years and who were at high genetic risk, after a latency of 5 years or more. These findings may support clinical decision-making related to radiation treatment, particularly among women for whom other modalities may be considered. The NHEJ GRS may be useful for individualizing both treatment and surveillance plans for young women who have received a first primary breast cancer diagnosis.


Liver cancer is the fastest increasing cancer in the United States and is one of the leading causes of cancer-related death in New York City (NYC), with wide disparities among neighborhoods. The purpose of this cross-sectional study was to describe liver cancer incidence by neighborhood and examine its association with risk factors. This information can inform preventive and treatment interventions.n = 6,407,022). Age-adjusted data on liver and intrahepatic bile duct cancer came from the New York State Cancer Registry (Publicly available data were collected on adult NYC residents (Registry  (2007–20p = 0.013); and with infection risk scores , particularly in the poorest neighborhoods . Some neighborhoods with high hepatitis rates do not have a proportionate number of hepatitis prevention services.Liver cancer incidence in NYC was strongly associated with neighborhood poverty after adjusting for race/ethnicity (β = 0.0217, High liver cancer incidence is strongly associated with infection risk factors in NYC. There are gaps in hepatitis prevention services like syringe exchange and vaccination that should be addressed. The role of alcohol and metabolic risk factors on liver cancer in NYC warrants further study. Cancer of the liver and intrahepatic bile ducts (liver cancer) is a public health problem in the United States (US). Since 1980, its nationwide incidence rate has tripled, and its mortality rate has doubled, outpacing the increase in any other cancer . Only 31In New York City (NYC), liver cancer is the fifth leading cause of cancer-related death among men and seventh among women . The 201New York City is a microcosm of the global population due to its unique demographics, high percentage of foreign-born inhabitants, and diversity of country of origin. Although a recent review examined racial/ethnic liver cancer disparities in the US , it has n = 6,407,022 per the 2010 US Census) at the neighborhood level, and defined neighborhood borders using NYC United Hospital Fund (UHF) codes. Originally, the UHF divided NYC into 42 distinct neighborhoods by combining adjoining zip codes areas with similar characteristics, meant to approximate NYC Community Planning Districts. To increase statistical power, these were later collapsed into 34 neighborhoods (Data were collected from the pool of adult (≥18 years) NYC residents , diabetes , and physical activity (% of adults who reported in the past 30 days: (1) exercising  and (2) walking/bicycling >10 blocks for transportation) were obtained via the CHS , injecting drug use (% of people who reported having used a needle to inject non-prescription drugs at least once), obesity  during 2014 were abstracted from the mandatory NYS hospital discharge abstract database , using tData on the availability of preventive services providing hepatitis B or C testing and treatment, hepatitis B vaccination, and syringe exchange facilities were collected from the NYC Health Map website , which lr ≤ 0.3, 0.3 < r < 0.5, r ≥ 5). Prevalence scores were multiplied by the correlation scores to obtain item scores, which were summed up to produce a risk score for each domain.Descriptive data are presented for the entire city, and each neighborhood in the form of tables and density maps prepared using ArcGIS Desktop . Predictors were weighted risk scores calculated for three domains of modifiable liver cancer risk factors: (1) metabolic ; (2) alcohol-related morbidity ; and (3) infections . Each continuous item was given an ordinal score based on tertiles, quartiles, or a specific cutoff. For each item, neighborhoods received a prevalence score from 1 to 3 based on increasing tertiles (quartiles 1–4 for hepatitis B). Hepatitis B vaccine coverage was reverse scored to reflect a protective effect. Due to the distribution, a cutoff of <1% and ≥1% was used to score injecting drug use prevalence category as 1 or 2. Each item was also assigned a correlation score from 1 to 3 based on the strength of its correlation with liver cancer incidence (Pearson’s I statistic)  . A sensin = 34) . All models met the assumptions for the specified Poisson distribution .The relationship between liver cancer incidence and each predictor was assessed in unadjusted and adjusted generalized linear regression models, with neighborhood as the unit of analysis  are female, 21% live in poverty, and 13% are uninsured. There was considerable variation in the distribution of demographic characteristics according to neighborhood , Fordham/Bronx Park (15), and Pelham/Throgs Neck (13.7)]; Manhattan , and Brooklyn [Sunset Park (16.7)] , even after adjustment for White race and Hispanic ethnicity . A higher proportion of foreign-born residents was correlated with higher rates of hepatitis B .Neighborhood-level data on poverty and Hispanic ethnicity were associated with high liver cancer incidence , followed by alcohol-related morbidity  (Table p = 0.052). We conducted formal testing by including interaction terms between poverty tertiles and each of the three risk scores , and observed lack of statistical interaction. When stratified by tertiles of poverty prevalence, infection score was most strongly associated with liver cancer incidence at the high poverty level . Similarly, infection risk score was most strongly associated with liver cancer incidence , especially at the high poverty level , in models that adjusted for metabolic and alcohol risk score.Among the three modifiable risk factor domains, infection was the strongest predictor of liver cancer incidence, with an expected increase of 5.3% in incidence when the infection risk score increased by 1 . Comparison of ordinary least squares regression and a spatial lag model found no meaningful effect of spatial autocorrelation on model estimates .The Moran’s The number of centers offering preventive services are 80, 89, and 28, respectively, for hepatitis B testing, treatment, and vaccination; 127 and 128, respectively, for hepatitis C testing and treatment; 23 syringe exchange programs (SEPs), with multiple additional distribution locations, and hundreds of Expanded Syringe Access Program locations throughout NYC. Their availability in relation to hepatitis burden is depicted in Figures To the best of our knowledge, this is the first study to describe liver cancer incidence and the distribution of its underlying risk factors at a neighborhood level in NYC. In addition, this is the first study to assess the availability of hepatitis prevention and treatment services in the context of disease burden.Results indicate that not only does NYC have higher rates of liver cancer incidence and mortality compared with NY State and the rest of the US but also large disparities exist among city neighborhoods, with incidence rates in some neighborhoods as high as those in China and West Africa . The mosRecent immigration patterns may also contribute toward the observed increase in hepatitis B. Between 2000 and 2011, NYC has seen a 4% increase in foreign-born residents . NeighboPreventive and treatment services for hepatitis are available throughout the city, but not all neighborhoods with high hepatitis rates have a proportionate number of required services. We observed that high hepatitis B rates were correlated with lower vaccine coverage and lower proportion of free vaccination centers. Lower insurance coverage was also strongly correlated with lower vaccine coverage. Hepatitis B vaccination can cost $120–$370 without insurance, plus consultation/professional administration fees. This is largely unaffordable for less affluent, uninsured people. Non-monetary factors such as having a vaccinated acquaintance, perceived risk of disease, perceived vaccine safety, and provider recommendation may also influence patients’ choice to receive the hepatitis B vaccine . TherefoWhile hepatitis C-related services were found to be more numerous, some neighborhoods appear to have fewer than 1 SEP per 1,000 hepatitis cases (Fordham–Bronx Park and Bedford–Stuyvesant–Crown Heights), while others (Coney Island and West Queens) have fewer than one hepatitis C testing and treatment centers per 100 hepatitis C cases. Although residents in poorer neighborhoods were more likely to get tested for hepatitis C, there is no information on how many of those who tested positive cleared the virus or received treatment. Without insurance, hepatitis C drugs for a 12-week course can cost between $39,600 and $94,500 , 31. Even = 34). Age, sex, and racial/ethnic diversity are other potential source of variation; however, due to the non-individual nature of the data and multiple race indicators, it was not possible to adjust for these variables. Alcohol risk scores were moderately correlated with metabolic and infection risk scores; however, results of statistical tolerance tests did not indicate a significant threat of multicollinearity on the model estimates. The NYC Health Map website does not provide data on private medical offices which may provide vaccinations and/or treatment whose information is not publicly available. Finally, we did not have data on homeless or incarcerated populations, who have an even higher risk of hepatitis C , Icahn School of Medicine at Mount Sinai. The protocol was approved (under exempt status) by the Institutional Review Board at the Icahn School of Medicine at Mount Sinai. A consent waiver was obtained since the data were de-identified, publicly available data.GK, ET, and NB had full access to all the data in the study and take responsibility for the integrity of the data and the accuracy of the data analysis. Study concept and design; study supervision: ET and NB. Acquisition, analysis, or interpretation of data: GK, NE, and SE. Drafting of the manuscript: GK, ET, NB, PP, JW, and MS. Critical revision of the manuscript for important intellectual content: GK, ET, NB, NE, PP, JW, MS, JL, and SE. Statistical analysis: GK.JW has received research grant support and served as a consultant for Gilead Sciences, Inc. JL is a consultant for Bayer, BMS, Eisai, Celsion, Eli Lilly and has active funding for research from Bayer, BMS, and Eisai. NB has active funding from Pfizer for research. All the other authors report no conflicts of interest.


BRCA variants have a high lifetime risk of developing breast and ovarian cancer. The aim of this study was to investigate the standard incidence ratios (SIR) for breast and ovarian cancer and standard mortality ratios (SMR) in a population-based cohort of women in Western Sweden, under surveillance and after risk reducing surgery. Women who tested positive for a BRCA variant between 1995–2016 (n = 489) were prospectively registered and followed up for cancer incidence, risk reducing surgery and mortality. The Swedish Cancer Register was used to compare breast and ovarian cancer incidence and mortality with and without risk reducing surgery for women with BRCA variants in comparison to women in the general population. SIR for breast cancer under surveillance until risk-reducing mastectomy (RRM) was 14.0 (95% CI 9.42–20.7) and decreased to 1.93 (95% CI 0.48–7.7) after RRM. The SIR for ovarian cancer was 124.6 (95% CI 59.4–261.3) under surveillance until risk reducing salpingo-oophorectomy (RRSO) and decreased to 13.5 (95% CI 4.34–41.8) after RRSO. The SMR under surveillance before any risk reducing surgery was 5.56 (95% 2.09–14.8) and after both RRM and RRSO 4.32 (95% CI 1.62–11.5). Women with cancer diagnoses from the pathology report after risk reducing surgery were excluded from the analyses. Risk reducing surgery reduced the incidence of breast and ovarian cancer in women with BRCA variants. However, overall mortality was significantly increased in comparison to the women in the general population and remained elevated even after risk reducing surgery. These findings warrant further research regarding additional measures for these women.Women with  BRCA1 and BRCA2 genes were discovered in the early 1990s [BRCA screening was implemented for women with hereditary breast and ovarian cancer. Subsequently, each oncogenetic clinic was comprised of a team of oncologists, clinical geneticists, clinical laboratory geneticists and oncology nurses that could provide genetic counseling and screening for patients and families with hereditary breast and ovarian cancer. The oncogenetic clinic at Sahlgrenska University Hospital in Gothenburg, Sweden was established in 1995 and currently serves Western Sweden with a population of 1.9 million inhabitants. Intriguingly, a novel recurrent founder variant in BRCA1 with origins in Western Sweden was discovered during the first years of BRCA screening at the Clinical Genetics laboratory [BRCA variants at the oncogenetic clinic to ensure that these women were given access to relevant clinical surveillance programs, and to be able to evaluate the long-term effects of measures, e.g. risk reducing surgery, performed within the cohort. Referral criteria for the assessment of hereditary breast and ovarian cancer has followed previous and current national healthcare programs (https://www.cancercentrum.se), and are in general similar to international guidelines (https://www.nccn.org).The boratory ,4. DurinBRCA variants were estimated. The risks for BRCA1-carriers were 72%  and 44% , and for BRCA2-carriers 69%  and 17% , respectively [BRCA showed a risk reduction of 90–100% of breast cancer after RRM and 72–88% for ovarian cancer [BRCA1 and 0.10% for women with BRCA2, and in another study there was a 20-year cumulative risk of peritoneal cancer of 3.5% [In a prospective cohort study with data from three international consortia, the cumulative risk for breast and ovarian cancer at 80 years of age for women with  95% CI, 5–79% andBRCA variants after risk reducing surgery in a defined geographic region in comparison with the general female population in the same region.Further, RRSO substantially reduced the increased mortality, both breast cancer–specific and ovarian cancer–specific mortality and all-cause mortality ,8,9. To BRCA1 variants and from 40 years of age for women with BRCA2 variants [National guidelines regarding hereditary breast and ovarian cancer recommends annual diagnostic imaging with breast magnetic resonance imaging (MRI) and breast ultrasound from ages 25 to 55 and mammography from ages 25 to 74 years. RRM has routinely been offered as a choice for those women who wanted a higher degree of risk reduction instead of surveillance. Regarding the ovaries, annual gynecological ultrasound was offered from 30 years of age. But more importantly, RRSO was recommended from 35 years of age for women with variants ,10. DuriBRCA variants. We evaluated cancer incidence and mortality among cancer-free women who underwent RRM and RRSO in comparison with baseline risks without risk reducing surgery and in comparison with the general female population.The aim of the present study was to assess the impact of risk reducing surgery on cancer incidence and mortality during a 20-year period (1995–2016) in a geographically defined population-based cohort of women in Western Sweden with BRCA between 1995–2016 at the regional oncogenetic clinic in Western Sweden and were given information about participating in a follow-up register to evaluate the long-term effects of surveillance and risk reducing surgery. In total, 489 women with BRCA variants were included in the register. The register contained information on date of DNA analysis, the identified BRCA variant, date of any previous malignancy, date of annual breast MRIs, ultrasounds, mammograms, date of annual gyneacological ultrasounds, date of risk reducing mastectomy, date of risk reducing oophorectomy, date and type of any cancer diagnosis after registration, date of death, and cause of death. Approval of the register was obtained from the Swedish Central Ethical Review Board , and all patients included in the register signed a consent form.The study population consisted of women that had tested positive for disease causing variants in BRCA variants during this 20-year period, either via BRCA screening or tested for a known familial variant. The analytical cohort consisted of the 253 women who were cancer-free when registered . The median age at the end of follow-up was 46.9 years of age . All instances of breast and ovarian cancer diagnosed after the DNA analysis, i.e., during the surveillance period, were classified as diagnosed before, at, or after risk reducing surgery. A diagnosis at risk reducing surgery was defined as diagnosed from the final pathology report or within 90 days after surgery. These cases were excluded from the analysis of SIR for breast and ovarian cancer and for the analyses of SMR.Four of the 253 women had already undergone a bilateral RRM before the time of DNA analysis, and therefore 249 women and 498 breasts were under surveillance after the time of DNA analysis. Fourteen of the 253 women had undergone a RRSO before the time of DNA analysis, and therefore 239 women with ovaries were under surveillance after the time of DNA analysis. When analyzing breast cancer incidence for women with breasts, breast years were calculated from the time of DNA analysis to the first of following events; diagnosis of breast cancer, RRM, death or end of surveillance period, i.e., 31 December 2016. Every breast was accounted for separately. For women who underwent RRM, breast years were calculated from the time of RRM to the first of following events; diagnosis of breast cancer, death or end of surveillance period, i.e., 31 December 2016. The SIR was defined as the observed number of breast cancers during these breast years divided by the expected number of cases, using incidence rates from the Swedish female population stratified for 5-year age groups  and calendar year.When analyzing ovarian cancer incidence for women with their ovaries intact, person years were calculated from the time of DNA analysis to the first event; ovarian cancer, RRSO, death or end of surveillance period. For women who underwent RRSO, person years were calculated from the time of RRSO to the first event; peritoneal cancer assessed to be of ovarian origin, death or end of surveillance period. SIR was defined as the observed number of peritoneal cancers of ovarian origin during these person years divided by the expected number of cases, using incidence rates from the Swedish female population stratified for 5-year age groups  and calendar period. Incidence rates for breast and ovarian cancer produced by the NORDCAN project  were useEvents of peritoneal cancer of unknown origin were not included in the analyses of SIR and SMR but were described in the results section. The reason for this was that the incidence rate of peritoneal cancer in the background population is uncertain.When analyzing the mortality rate for all women in the cohort, person years were calculated from the time of DNA analysis to death or end of surveillance period. When analyzing the mortality rate for women before any risk reducing surgery, person years were calculated from the time of DNA analysis to bilateral RRM or RRSO or death or end of surveillance period. When analyzing the mortality rate for women after bilateral RRM or after RRSO, person years were calculated from bilateral RRM or RRSO to death or end of surveillance period. Lastly, when analyzing the mortality rate for women after both bilateral RRM and RRSO, person years were calculated from the last risk reducing surgery event to death or end of surveillance period. The SMR was defined as the observed number of deaths during these person years divided by the expected number of deaths, using rates from the Swedish female population stratified for 5-year age groups  and calendar year.The stset, stsplit and strate macros in STATA 15.1 were used for the incidence analyses .In total, 31 breast cancers were diagnosed for 249 women who were originally cancer-free when registered, of which 25/31 were diagnosed during surveillance, 4/31 were diagnosed in the final pathology report from RRM and 2/31 cancers were diagnosed after the women underwent a RRM .BRCA1, 4 were associated to BRCA2. Also, 25 breast cancers were observed in women under surveillance, while 1.79 were the number expected, resulting in a SIR of 14.0  and the median age for RRM was 42.3 years . Thirty women had 31 breast cancer diagnoses; 27 cancers were associated to Eighty women had a bilateral RRM during surveillance, whereas 24 women had a unilateral RRM after being diagnosed with breast cancer during surveillance, i.e., in total 184 breasts. Two cancers were observed after risk reducing surgery while the expected number was 1.0, SIR 1.93 (95% 0.48–7.70). The youngest woman was diagnosed at age interval 35–39, and one woman at age interval 45–49.BRCA1 variants. Seven of the 14 ovarian cancers were observed during surveillance, while the expected number was 0.06, with a SIR of 124.57 . Also, 3/4 of the deceased women were in the 45–49 age interval, and breast cancer was the cause of death for only one of them. In total, 136 women had a RRSO and during follow-up 11 deaths were observed, while the expected number was 3.7, SMR 2.99 (95% CI 1.66–5.40). The youngest patient to die was in the 40–44 age interval and four in the 45–49 age interval. Five of the 11 deaths after RRSO were due to ovarian or peritoneal cancer.Among 62 women that had undergone both RRM and RRSO, 4 deaths occurred, 3 from peritoneal cancer of ovarian origin and 1 with a non-malignant cause of death, SMR 4.32 (95% CI 1.62–11.5). Three of 62 women had an ovarian cancer diagnosis from the final pathology report and as previously mentioned, these were excluded from the analyses of SIR and SMR. These women did not die during the surveillance period.BRCA variants, primarily registered at the time of genetic testing and thereafter prospectively followed regarding cancer incidence, risk reducing surgery and death between 1995–2016. In concordance with previous studies, we observed a reduction in breast and ovarian cancer incidence after RRM and RRSO as well as mortality [BRCA variants were still significantly increased compared to age matched-women in the general population, even after risk reducing surgery, SMR 4.32 . As far as we know, a persistent increase in SMR even after risk reducing surgery has not been previously shown in a study with a population-based cohort from a geographically defined region. These findings emphasize the need for further research to identify additional complementary measures than risk reducing surgery for high-risk women with BRCA variants.This population-based follow-up study was restricted to a defined geographic region, Western Sweden, with a cohort of women with ortality ,9. HowevFor example, a more precise and individualized risk evaluation combining genetic screening for polygenic risk score in combination with hormonal factors and family pedigree could influence the woman to take preventive measures; engaging in surveillance programs or having risk reducing surgery, in time . FurtherBRCA1 and 40 years of age for BRCA2. These recommendations were thus adequate in relation to the ages of cancer diagnoses in our cohort.In our cohort, we observed two cases of breast cancer among 80 women who had a bilateral RRM. Furthermore, we observed three cases of peritoneal cancers of ovarian origin after RRSO among 136 women. This is in line with studies showing no or small remaining risks for cancer  after riBRCA1 variants and only 40 had BRCA2 variants, which affects the generalizability as BRCA1 is associated with a higher risk of both breast and ovarian cancer. Further, there is in the literature evidence for variant genotype–phenotype correlations in BRCA genes [BRCA1 variants carried the founder variant c.3171ins5 with an origin in Western Sweden, but if there is a correlation between this very variant and the breast and ovarian cancer ratios, is not known. BRCA variants may be associated with other cancers such as pancreas and colon cancer, in particular BRCA2 variants are associated to a relatively low increase in risk of pancreas cancer [In this study cohort of 253 women, 213 had s cancer . Those tBRCA variant were asked to participate in a prospective follow-up register study. There were only a few lacking consents and the genetic counselors did not perceive the lacking consents to be related to whether the patient had cancer or risk reducing surgery. The high degree of accepting registration and surveillance in our cohort may depend on the fact that healthcare in Sweden is mainly financed by taxes, access to healthcare is subsidized or free for the patient, and the healthcare region is responsible for funding and providing healthcare services, including this type of surveillance program, risk reducing surgery and cancer treatment.All women in the region of Western Sweden with a BRCA variants in a defined geographic region during the period 1995–2016. Follow up was complete and the medical exams during surveillance as well as risk reducing surgery were well described in the register. Limitations of the study was the limited number of person years of follow-up and limited time of follow-up due to the majority of the women entering the study during the latter part of the inclusion period, at which time genetic testing gradually became more common. This resulted in wide confidence intervals for the estimates.A strength of this study was the prospectively-followed primarily registered cohort of 489 women with BRCA variants had a remaining increased mortality in comparison to women in the general population, even after risk reducing surgery. This persistent increase in mortality calls for further measures than just risk reducing surgery in the risk management of women with BRCA variants.In this population-based study, women with


Breast cancer patients who have not previously attended mammography screening may be more likely to discontinue adjuvant hormone therapy and therefore have a worse disease prognosis.We conducted a population-based cohort study using data from Stockholm Mammography Screening Program, Stockholm-Gotland Breast Cancer Register, Swedish Prescribed Drug Register, and Swedish Cause of Death Register. Women in Stockholm who were diagnosed with breast cancer between 2001 and 2008 were followed until December 31, 2015. Non-participants of mammography screening were defined as women who, prior to their breast cancer diagnosis, were invited for mammography screening but did not attend.Of the 5098 eligible breast cancer patients, 4156 were defined as screening participants and 942 as non-participants. Compared with mammography screening participants, non-participants were more likely to discontinue adjuvant hormone therapy, with an adjusted hazard ratio (HR) of 1.30 . Breast cancer patients not participating in mammography screening were also more likely to have worse disease-free survival, even after adjusting for tumor characteristics and other covariates .Targeted interventions to prevent discontinuation of adjuvant hormone therapy are needed to improve breast cancer outcomes among women not attending mammography screening.The online version of this article (10.1186/s12916-019-1252-6) contains supplementary material, which is available to authorized users. Breast cancer is the most frequently diagnosed cancer and the second leading cause of cancer deaths among women, accounting for 15% of total cancer deaths worldwide and 13% of total cancer deaths in Sweden in 2012 –9.Adjuvant hormone therapy—using tamoxifen and aromatase inhibitors—has been reported to lower the risk of breast cancer recurrence by over 30% . DespitePrevious studies have shown that barriers to mammography screening adherence may also prevent patients from adhering to subsequent adjuvant hormone therapy , 19–21. This study was approved by the Regional Ethical Review Board in Stockholm, Sweden .Since 1989, The Stockholm Mammography Screening Program invited all women in Stockholm aged 50 to 69 years for mammography screening at 24-month intervals, and since 2005, women aged 40 to 49 years have been invited at 18-month intervals –24. The n = 680) and with distant metastasis at cancer diagnosis (n = 77), leaving a total of 5098 patients for the final analysis  or aromatase inhibitors (ATC codes L02BG).Non-participants of mammography screening were defined as women who were invited for mammography screening 2 years  before their breast cancer diagnosis but did not attend. The above cutoffs were used given that this is the longest screening interval applied in the Stockholm Mammography Screening Program throughout the study period –24.Information on age at diagnosis, menopausal status at diagnosis, family history of breast cancer, tumor size, lymph node involvement, estrogen receptor status, progesterone receptor status, HER2 status, and tumor grade was retrieved from the Stockholm-Gotland Breast Cancer Register. Information on country of birth and marital status at diagnosis was retrieved from the Total Population Register . InformaDiscontinuation of adjuvant hormone therapy was defined as having any interval between two consecutive dispenses of greater than 180 days during the follow-up , 31. In Discontinuation was defined by monitoring patients from the first prescription of tamoxifen or aromatase inhibitors until death, local recurrence, distant metastasis, contralateral breast cancer, emigration, end of study period , or completion of a 5-year treatment, whichever came first. Time to discontinuation was calculated as the interval between the first and last prescription, added to the number of days of supply from the last prescription.Patients were monitored from the date of cancer diagnosis until death, emigration, or end of study period , whichever came first. Information on local recurrence, distant metastasis, and contralateral breast cancer was retrieved from the Stockholm-Gotland Breast Cancer Registry. Information on date of emigration was retrieved from the Swedish Emigration Registry. Information on cause of death was retrieved from the Swedish Cause of Death Registry.Time to breast cancer event was defined as the time from cancer diagnosis to local recurrence, distant metastasis, contralateral breast cancer (> 3 months after the primary breast cancer), or death from breast cancer, whichever came first .p value < 0.05 were included in the multivariable analyses. Kaplan-Meier analysis and Cox regression analysis were used to compare differences in discontinuation of adjuvant hormone therapy and breast cancer events among screening participants versus non-participants. The proportionality assumption for running a Cox model was checked using the Schoenfeld residual test, with no model violation observed.Chi-square tests (or Fisher’s exact test if the expected cell frequencies were less than 10) were used to compare differences in baseline and tumor characteristics among screening participants versus non-participants. Only characteristics with a We repeated our analyses by comparing screening non-participants with subgroups of screening participants: patients with screen-detected cancers (cancer detected after a positive screening result) and interval cancers (cancer detected after a negative screening mammography but before the next scheduled examination), given that these are both diverse groups with different tumor characteristics and breast cancer outcomes.We also repeated our analysis of screening non-participation and breast cancer events using a competing risk regression model, accounting for non-breast cancer deaths as a competing event.p < 0.05.All statistical analyses were two-sided and performed using SAS version 9.4  or Stata version 13.0 . Statistical significance was determined at Table Figure Figure Consistently higher discontinuation rates were found when comparing non-participants with subgroups of screening participants: an adjusted HR of 1.31  when compared to screen-detected cancers and 1.25  when compared to interval cancers  and 95% confidence intervals (95% CIs) for discontinuation of adjuvant hormone therapy and breast cancer events derived from competing risk regression models. Discontinuation of adjuvant hormone therapy and breast cancer events in women diagnosed with breast cancer in Stockholm, Sweden, 2001–2008. (A) Screening non-participants vs participants; (B) screening non-participants vs participants diagnosed with screen-detected cancers; (C) screening non-participants vs participants diagnosed with interval cancers. Hazard ratio (HR) and 95% confidence intervals (95% CIs) were derived from competing risk regression models. (DOCX 13 kb)


Denosumab inhibits the receptor activator of nuclear factor κB (RANK) pathway and is used to treat osteoporosis. Emerging evidence suggests RANK-blockade may play a role in mammary tumourigenesis. Thus, we undertook a population-based study of denosumab use and breast cancer risk in a large cohort of postmenopausal women.We included women 67+ years with prior bisphosphonate use who filled a first prescription for denosumab. They were matched on age, date, cumulative prior use of and time since last use of a bisphosphonate to women with no history of denosumab. Cox proportional hazards was used to estimate the hazard ratio (HR) of breast cancer with denosumab use.P-trend = 0.15).A total of 100,368 women were included in the analysis with 1271 incident breast cancer events. Denosumab use was associated with a 13% decreased breast cancer risk . There was no relationship between increasing number of denosumab doses and breast cancer risk (These findings suggest a potential protective effect of ever denosumab use on breast cancer risk in a cohort of older women previously treated with bisphosphonates. In contrast, binding of RANKL either pharmacologically or by osteoprotegerin (OPG --the endogenous decoy receptor for RANKL) inhibits the RANKL-mediated signaling pathway, consequently inhibiting bone resorption and maintaining bone density. A large trial of over 7000 older women with osteoporosis demonstrated that denosumab significantly reduced the risk of fractures.3 Based on those findings, denosumab was added to the Ontario Drug Benefits formulary in 2012 for the treatment of osteoporosis under restricted criteria that requires prior exposure to, or hypersensitivity following oral bisphosphonate use.4 It is estimated that 1% of older women in Ontario have initiated this drug annually since its addition to the provincial drug formulary.5 The typical schedule for denosumab is one subcutaneous injection every 6 months.The drug denosumab is an anti-RANKL monoclonal antibody which is used to treat osteoporosis and prevent skeletal damage caused by breast cancer metastases.10 Furthermore, preliminary findings from the adjuvant denosumab in breast cancer (ABCSG 18) double-blind, placebo-controlled trial showed improved disease-free survival among the women randomized to denosumab injected subcutaneously twice a year.12 The impact of denosumab on the incidence of a second primary cancer has yet to be reported. Targeting of RANK-signaling may be particularly relevant for women at a high risk of developing breast cancer attributed to an inherited mutation in BRCA1.10There is emerging evidence that progesterone-mediated up-regulation of the RANK/RANKL also plays a critical role in mammary gland epithelial cell proliferation, in mammary stem cell expansion, and in mammary carcinogenesis.13 Given the suboptimal uptake of the current prevention options, it is important to identify novel and highly effective therapeutic cancer prevention strategies.14 Given the seminal preclinical evidence supporting a role of aberrant RANK-signaling in the development of breast cancer, it is of interest to evaluate whether denosumab with a relatively safe toxicity profile, is a potential candidate.15 To our knowledge, there are no studies that have specifically evaluated the relationship between denosumab use and breast cancer risk. The objective of the current study was to utilize large healthcare administrative databases to evaluate the incidence of breast cancer in a large cohort of postmenopausal women following denosumab initiation.The current landscape of chemoprevention for women at high-risk of developing breast cancer consists of either selective estrogen receptor modulators  or aromatase inhibitors and is dependent on menopausal status.16 to identify invasive breast cancer and cancer history; the Canadian Institute for Health Information Hospital Discharge Abstract Database (CIHI-DAD/) for information regarding hospital admissions; the National Ambulatory Care Reporting System (NACRS) for emergency department visits and day surgeries; the Ontario Drug Benefit (ODB) database for prescription drug claims records as all residents of Ontario aged 65 years and older are eligible for provincial drug coverage through the ODB;18 and the Ontario Health Insurance Plan (OHIP) for information about physician service claims including mammography history. These datasets were linked using unique encoded identifiers and analyzed at the Institute for Clinical Evaluative Sciences (ICES).We conducted a population-based matched cohort study using healthcare administrative databases in Ontario, Canada. The databases contain records for all individuals eligible for the province’s universal health coverage. The databases included in this analysis were: the Registered Persons Database files for demographic information ; the Ontario Cancer Registry (OCR)This study was approved by the Research Ethics Board of the Sunnybrook Health Sciences Centre and Women’s College Hospital.21 we restricted inclusion to women with prior oral bisphosphonate use to isolate effects of denosumab. Three oral bisphosphonates are approved for osteoporosis in Canada , and have been available through the ODB program since 1996.22 Bisphosphonate use prior to the date of study entry  was obtained for each subject and the cumulative use and time since most recent use was calculated. Women were excluded if they were over the age of 85 at index date, ineligible for OHIP at any point in the two years preceding the index date, had no ODB claims in the two year preceding the index date, had a history of any cancer  at any point prior to the index date, had a history of any conditions that would impact bone quality at any point prior to the index date , had a death date preceding/on the index date and had a history of living in a long-term care setting.Given that eligibility for ODB coverage of denosumab is predominantly restricted to women with a history of oral bisphosphonate use, and breast cancer risk is suggested to be lower among women with osteoporosis and also following bisphosphonate exposure;We identified all women aged 67 or more years who received a first prescription of denosumab between 29 February 2012 (first date on ODB formulary) and 30 April 2016. The index date was the first date of dispensation of denosumab. The total number of denosumab prescriptions dispensed at least four months from each other ( ≥ 120 days) were then categorized as 1–2 doses, 3–4 doses, and ≥ 5 doses. All women meeting our inclusion criteria with oral bisphosphonate exposure since 1996 were assigned a random index date based on the distribution of index dates among the eligible exposed subjects.23 as well as history of pathologic or other fractures, documentation of osteoporosis, and prior use of other drugs that may impact bone health .We also collected information on the following covariates: age, resident location  determined by linking postal codes to census data, income status based on neighborhood income quintile, number of primary care visits, emergency department (ED) visits or acute care hospitalizations in the previous year, screening mammogram in the two years prior to the index date and comorbidity using the John Hopkins aggregated diagnosis groups (ADG) score in the two years prior to the index date,All women aged 67 or more years with oral bisphosphonate exposure since 1996 were assigned a random index date based on the distribution of index dates among the eligible exposed subjects; this ensured that exposed women could serve unexposed time prior to denosumab initiation. We conducted 3:1 matching of unexposed to exposed subjects. Subjects were matched on age ( ± 2 years), index date ( ± 1 year), cumulative use of bisphosphonates ( ± 1 year), time since last use of a bisphosphonate ( ± 1 year) and propensity score. Propensity scores were generated using variables for urban vs. rural residence, historic ED and inpatient visits, ADGs, history of fracture, and historic use of any anti-estrogen therapy, estrogen therapy, or aromatase inhibitors. We conducted propensity score calliper matching using a calliper of 0.2 times the standard deviation of the propensity score.The primary outcome of interest was a diagnosis of incident invasive breast cancer documented in OCR in the follow-up period. Incident breast cancers included any malignant neoplasms of the breast (ICD-10 any C50). Benign neoplasms of the breast (ICD-10 any D24), carcinomas in situ of the breast (ICD-10 any D05) and neoplasms of uncertain behavior of the breast (ICD-10 D485 or D486) were not included.24 Cox proportional hazards models, stratified on matched pairs was used to estimate the adjusted hazard ratio (HR) and 95% confidence intervals (CI) for denosumab exposure, as well as increasing dose and time since last bisphosphonate use, and the risk of breast cancer. Women were followed from their index date to either: (1) an incident breast cancer diagnosis, (2) other cancer diagnosis (excluding non-melanoma skin cancer), (3) death, or (4) end of follow-up (31 August 2017), whichever occurred first. Unexposed subjects who eventually received denosumab in the follow-up were also censored .Baseline descriptive characteristics of the two groups were compared using standardized differences. A standardized difference of <0.10 was used to determine comparability between the groups for each covariate of interest.All statistical analyses were performed using SAS software version 9.3.A total of 100,368 women with a history of bisphosphonate use were included in the final analysis. Of these, 25,092 women were denosumab users and 75,276 were matched non-users. Prior to matching, users and non-users of denosumab differed with respect to age, residential location, bisphosphonate use, history of fall-related injuries, fractures, primary care visits, emergency department visits and acute care hospitalizations in the past year, as well as, total aggregated diagnosis groups. However, following matching for age, index date, bisphosphonate use and propensity score, the two groups of women were similar with respect to all the baseline characteristics (Table P – log rank = 0.04), Figure P – trend = 0.15) or time since last use of bisphosphonates (P – interaction = 0.52)  cases diagnosed among the denosumab users and 986 (1.3%) cases among the non-users (Table The characteristics of the breast cancer patients are summarized in Table P = 0.04)(data not shown).In a sensitivity analysis, the inclusion cohort was re-matched such that the index date for denosumab users was 6 months following their initial injection to allow for a 6 month lag period between denosumb use and breast cancer risk. The findings did not change considerably. For example, the HR comparing users vs. non-users was 0.86 . There was no dose-response relationship and the association did not vary by time since last bisphosphonate use. Except for nodal involvement, pathologic, and histologic characteristics of the breast tumors did not vary by history of denosumab exposure. Although these findings suggest a protective effect of denosumab exposure, they should be interpreted with caution given the relatively short duration of follow-up in this analysis as well as the median age of the population (~76 years). It is of interest to confirm our findings in other studies and to further evaluate the relationship between denosumab and breast cancer in younger patients and with additional years of follow-up.25 On the other hand, we cannot rule-out some residual confounding related to denosumab initiation. For example, women treated with denosumab (a new drug to market) may be healthier in terms of diet, alcohol consumption and exercise – aware of the new pharmacological option and broaching the discussion with their physician, or treated by attentive physicians encouraging preventive health behaviors.Our cohort consisted of women from the general population with a history of bisphosphonate use, and thus, are not representative of women in the general population. Although we found little evidence of a dose-response association between denosumab exposure and subsequent breast cancer risk, our findings with any denosumab use do not preclude a potential chemoprevention role of this drug among women without a history of osteoporosis including those under the age of 67. Indeed, our results suggests that a short-course of denosumab has the potential to offer long-term protection against breast cancer which is analogous to the cancer protective effects conferred by a later age at menarche and breastfeeding, intrinsic, and transient exposures that significantly reduce the risk of breast cancer.BRCA1 mutation. Pre-clinical findings from various seminal publications have collectively elucidated a pivotal role of the RANK-pathway in brca1 mammary carcinogenesis.27 Specifically, Nolan et al., demonstrated that RANKL inhibition resulted in a significant delay in mammary tumor onset and incidence in a brca1 deficient mouse model, and furthermore, that treatment of premenopausal women with denosumab resulted in a substantial reduction in breast epithelial cellular proliferation based on Ki67 expression27 and confirmed by an independent research group.26 These findings are of particular relevance for women with a BRCA1 mutation given their high lifetime risk of developing breast cancer, the very limited data regarding tamoxifen use for primary prevention, along with the suboptimal uptake of tamoxifen since most BRCA mutation carriers opt for yearly screening with MRI.28 Randomized trials or observational intervention trials in this specific population are highly anticipated. We did not have information on family history or BRCA mutation status, and thus, were not able to assess risk in these subgroups.Despite the inclusion of an older population in the current study, one cannot preclude a potential breast cancer protective effect of denosumab (or other RANKL inhibitors) in younger high-risk populations, particularly among women with an inherited 29 Intravenous bisphosphonates are also prescribed to breast (and other) cancer patients to prevent treatment-induced skeletal complications including bone loss and bone metastases.30 Evidence from earlier, epidemiologic studies suggested a possible reduction in breast cancer risk among postmenopausal women who used bisphosphonates;21 however, a results from a recent prospective cohort of 64,438 French postmenopausal women and 2407 incident cases, reported no significant association between bisphosphonate use and breast cancer risk .31The prevention and treatment of postmenopausal osteoporosis have historically included the use of bisphosphonates, a class of drugs that inhibits osteoclast-mediated bone resorption.32 Given that more than half of the breast cancers in Canada are diagnosed prior to age 69, we did not capture the full population of interest. The women included in the current analysis were limited to older (on average 76 years of age) women with a history of bisphosphonate use, and likely not representative of the larger number of women at risk of developing breast cancer. Although we did not have information on various breast cancer risk factors including family history, we were able to demonstrate that both groups of women were similar with respect to screening . This was purely attributed to the fairly recent introduction of denosumab for the treatment of osteoporosis. We wanted to ensure accurate information regarding prescribed denosumab use, and thus, only included women who were 67 years of age or older who had at least 2 prior years of coverage under the ODB program. However, it should be noted that denosumab for the treatment of osteoporosis was not added to the Ontario provincial formulary until February 2012 and is only provided in special circumstances. This may have resulted in some misclassification including under-capturing if patients received drug coverage through other mechanisms .Despite these limitations, our study had several strengths, in particular, the use of large provincial administrative datasets, allowing for well-powered analyses and matching on relevant confounders. Our exposed and unexposed groups were similar with respect to most demographic characteristics, prior history of bisphosphonate use and other medications that may impact bone health , comorbidities, as well as, health care utilization patterns. Comprehensive ODB drug data permits capture of oral bisphosphonate therapy since 1996, matching on age and calendar time helps control for changes in osteoporosis management and therapy over time.In conclusion, we found a small inverse relationship between denosumab exposure and breast cancer incidence in this large population-based study of older women residing in Ontario, Canada. To our knowledge, this represents the first report of denosumab use and subsequent breast cancer risk. Further studies with a longer follow-up period, as well as the inclusion of younger women or cohorts of high-risk women, are necessary to delineate the role of RANK-inhibition in the prevention of breast cancer. It is of importance to establish whether denosumab, administered subcutaneously as a semi-annual injection with a safe toxicity profile, has the potential to be used in the primary prevention setting.


We examined how breast cancer risk factors are associated with MD area (cmWe conducted a cohort study of 31 782 Swedish women ages 40–70 years at time of baseline mammogram. Lifestyle and reproductive risk factors were assessed by a web-based questionnaire. MD was measured as dense area using the STRATUS method (mean over the left and right breast). Linear regression analyses with adjustments for age, body mass index (BMI), and menopausal status at baseline were performed to assess the association between breast cancer risk factors and mean baseline MD. To investigate mean MD change across age, linear regression analyses with adjustments for age, BMI, menopausal status, and age at last mammogram were performed. All tests of statistical significance were two-sided.2. BMI and physical activity were statistically significantly associated with MD change. Lean women (BMI <20 kg/m2) had a mean MD change of −1.13 cm2 per year  compared with −0.46 cm2 per year  for women with BMI 30 or higher. The annual MD change was −0.4 cm2 larger in women who were very physically active compared with less physically active women.Except for oral contraceptive use, established lifestyle and reproductive risk factors for breast cancer were associated with baseline mean MD. The overall average annual MD change was −1.0 cmOur results indicate that all risk factors for breast cancer, except oral contraceptive use, are associated with baseline MD but that only age, BMI, and physical activity are determinants of MD change. Mammographic density (MD) reflects the radiologically dense part of a mammogram that consists of epithelial tissue and stroma that appear bright on a mammogram, whereas fat tissue appears dark . MD is oMost studies regarding the associations of established risk factors for breast cancer and MD have involved only a single mammographic examination. It has been shown that older age, more children, early pregnancy, postmenopausal status, and elevated BMI are associated with lower MD . In contIn a longitudinal study by Boyd et al. , the aveThe KARMA cohort is a population-based prospective screening cohort initiated in January 2011 and comprises women attending mammography screening or clinical mammography at four hospitals in Sweden . Women w2). STRATUS is a fully automated tool developed to analyze digital and analogue images using an algorithm that measures density on all types of images regardless of vendor , smoking status , alcohol consumption , physical activity , age at first birth , number of births , breast-feeding among parous women , time since last birth , menopause status , age at menarche (<13 or ≥13 years), contraceptive use , MHT use , and family history of breast cancer . Women reporting no natural menstruation over the past 12 months before the study entry or no menstruation due to oophorectomy were considered postmenopausal. Women with missing information on menstruation status or having no menstruation due to gynecological surgeries other than oophorectomy were considered premenopausal if they were age 50 years or younger and postmenopausal if older than 50 years.Participants completed a detailed web-based questionnaire approximately 3 months from date of the baseline mammogram. Lifestyle and reproductive factors were categorizes as: age at baseline , BMI  and the mean difference between each category and the reference. To enhance interpretability, we used the fitted regression models to estimate the mean MD for each risk factor. Technically, this was accomplished with regression standardization .To assess how mean MD changed across age, we fitted local polynomial regression curves using baseline mean MD and age at baseline as the dependent and independent variables, respectively. We fitted one curve for each level of the established breast cancer risk factors listed above. The obtained local polynomial regression curves enables a qualitative  assessment of how mean MD changes as a function of the established breast cancer risk factors, without making strong parametric assumptions. A disadvantage of these curves is that they provide no quantitative measures for testing whether mean MD change differs across levels of established breast cancer risk factors. Another disadvantage is that these curves discard large parts of the data by using only baseline measures of mean dense area.2 per year during follow-up for that woman. Second, the estimated dense area changes (slopes) were regressed on the lifestyle, reproductive, and established breast cancer risk factors listed above, using one linear regression model for each factor. This model was adjusted for age at first and last mammograms, baseline BMI, and baseline menopausal status. Adjustments for age at first and last mammograms are necessary to avoid confounding by age during follow-up, because the age at which the mammograms were taken is strongly related to the estimated dense area change and may also be related to several of the risk factors for breast cancer. Interaction analyses were performed to determine whether determinants of MD change differ by menopausal status.We carried out a more elaborate density change analysis in two steps. First, one linear regression model was fitted for each woman, regressing all her observed dense area measures during follow-up on her attained ages at these measures. The obtained slopes from these regressions quantify the Woman-specific dense area changes. For instance, a slope equal to −1.5 for a particular woman indicates that the dense area decreased with an average 1.5 cmLastly, we performed a sensitivity analysis adjusting for both baseline BMI and BMI change among 6427 women for which repeated BMI measures were available. That is, for each woman we have regressed her observed BMI measures during follow-up on her attained ages at these measures. The obtained slopes from these regressions quantify the Woman-specific BMI changes.All statistical analyses were two-sided and were performed using R version 3.4.1. Statistical significance was measured at significance level 0.05.2), 31.1% were considered overweight, and 11.6% were obese (≥30 kg/m2)  . Almost 0 kg/m2) . An age 0 kg/m2)  women. HPhysically active, having the first child early in life, having many children, breast-feeding longer than 6 months, having an early menarche, and having the last child 10 years or more years ago were associated with lower MD at baseline . Lastly,2. The overall average annual MD change was –1.0 cm2) had a mean MD change of –1.13 cm2/y (95% CI = –1.25 to –1.02) compared with –0.46 cm2/y (95% CI = –0.57 to –0.35) for women with a BMI of 30 or more kg/m2  (2). In agreement with our findings, they also showed that although age at first birth and number of children are associated with baseline percent MD, these factors do not seem to be associated with MD change over time  . In anotP = .01)  found nover time .BMI has been hypothesized to be associated with breast cancer risk through several hormonal-related mechanisms, which may also be relevant to MD change. Increased risk of breast cancer among overweight and obese women may be explained by a higher rate of conversion of androgenic precursors to estrogens through the peripheral aromatization in adipose tissue ,36. Estr = 722), Conroy et al. ( = 129), the frequency of participating in exercise for fitness or recreation was not associated with change in percent or area MD ( = 413) found no association between physical activity and change in percent MD ; Department of Oncology, Södersjukhuset, Stockholm, Sweden (PH).The authors declare that they have no conflict of interest.The authors thank all the participants in the KARMA study and personnel for their devoted work during data collection. They also would like to acknowledge Jose Tapia for helping in data management.SA participated in the study design, performed the statistical analyses, interpreted the results with support from AS, and drafted the manuscript. AS, ME, MG, KC, and PH participated in the study design and interpretation of the results and helped draft the manuscript. All authors read and approved the final manuscript.The study was approved by the ethical review board at Karolinska Institutet. Informed consent was obtained from all individual participants included in the study. All experiments comply with the current Swedish laws.Supplementary DataClick here for additional data file.