Abstract
Purpose
Nipple-sparing mastectomy, which may improve cosmesis, body image, and sexual function in comparison to non-nipple-sparing mastectomy, is increasingly used to treat early-stage breast cancer; however, long-term survival data are lacking. We evaluated survival after nipple-sparing mastectomy versus non-nipple-sparing mastectomy in a population-based cancer registry.
Methods
We conducted an observational study using the California Cancer Registry, considering all stage 0–III breast cancers diagnosed in California from 1988 to 2013. We compared breast cancer-specific and overall survival time after nipple-sparing versus non-nipple-sparing mastectomy, using multivariable analysis.
Results
Among 157,592 stage 0–III female breast cancer patients treated with unilateral mastectomy from 1988–2013, 993 (0.6 %) were reported as having nipple-sparing and 156,599 (99.4 %) non-nipple-sparing mastectomies; median follow-up was 7.9 years. The proportion of mastectomies that were nipple-sparing increased over time (1988, 0.2 %; 2013, 5.1 %) and with neighborhood socioeconomic status, and decreased with age and stage. On multivariable analysis, nipple-sparing mastectomy was associated with a lower risk of breast cancer-specific mortality compared to non-nipple-sparing mastectomy [hazard ratio (HR) 0.71, 95 % confidence interval (CI) 0.51–0.98]. However, when restricting to diagnoses 1996 or later and adjusting for a larger set of covariates, risk was attenuated (HR 0.86, 95 % CI 0.52–1.42).
Conclusions
Among California breast cancer patients diagnosed from 1988–2013, nipple-sparing mastectomy was not associated with worse survival than non-nipple-sparing mastectomy. These results may inform the decisions of patients and doctors deliberating between these surgical approaches for breast cancer treatment.
Keywords: Breast cancer, Mastectomy, Skin-sparing, Nipple-sparing, Survival
Introduction
Despite randomized clinical trials demonstrating equivalent survival after breast conserving therapy versus mastectomy [1], use of mastectomy (specifically, contralateral prophylactic mastectomy) has risen recently [2]. This coincided with increased uptake of genetic testing for cancer risk assessment [3, 4], and with reports that prophylactic mastectomy reduces breast cancer risk among women with an inherited BRCA1/2 mutation [5]. Given evidence that mastectomy rates are rising, interest has grown in less invasive procedures such as nipple-sparing mastectomy (NSM) [6]. Compared to non-nipple-sparing mastectomy (non-NSM), NSM may improve cosmesis, body image, and sexual function [7]. However, concerns remain about NSM’s safety with regard to breast cancer recurrence and survival. Randomized clinical trials do not exist and are unlikely to be initiated, and existing observational studies were limited to single centers or short follow-up time. We took advantage of the large population-based California Cancer Registry (CCR) to compare survival of stage 0–III female breast cancer patients treated with NSM versus non-NSM from 1988 to 2013.
Methods
The study population consisted of all female California residents diagnosed with a first primary breast cancer (International Classification of Disease for Oncology, 3rd Edition, site codes C50.0–50.9 and histologic codes: 8000, 8010, 8020, 8022, 8050, 8140, 8201–8230, 8255, 8260, 8401, 8453, 8480–8525, and 8575), of American Joint Commission on Cancer stages 0–III, from January 1, 1988 to December 31, 2013. The analysis was overseen by the Institutional Review Board of the Cancer Prevention Institute of California. We obtained CCR information regarding patient and tumor characteristics, initial treatment course and patient vital status through December 31, 2013. We used an established measure of neighborhood socioeconomic status (SES) based on patients’ residence when diagnosed [8]. An initial surgical procedure of subcutaneous mastectomy, also called nipple-sparing mastectomy, was coded as NSM. Procedures of total (simple) mastectomy, modified radical mastectomy, radical mastectomy, or extended radical mastectomy (all without removal of uninvolved contralateral breast) and mastectomy NOS were coded as non-NSM. Survival time was measured in days from diagnosis to death. We used Cox proportional hazards regression to model associations with overall and breast cancer-specific mortality. Minimally adjusted models were stratified by stage and adjusted for age. Fully adjusted models were stratified by stage and histology; adjusted for age, race/ethnicity, tumor size, lymph node involvement, adjuvant chemotherapy and/or radiation, neighborhood SES, marital status, hospital characteristics (SES composition of patients and National Cancer Institute-designated cancer center status), and diagnosis year; and adjusted for clustering by hospital. In secondary analyses limited to diagnoses in 1996 or later, for which more covariates were available, models were additionally adjusted for grade, estrogen receptor (ER)/progesterone receptor (PR) status, and insurance status. We tested the proportional hazards assumption for each covariate using correlation tests of time versus scaled Schoenfeld residuals. The assumption was violated for stage and histology; thus, we conducted stratified Cox regression models allowing the baseline hazard to vary by these variables. We used SAS version 9.4 for all analyses.
Results
A total of 547,893 women were diagnosed with a first primary breast cancer in California from 1988 to 2013. Patients were excluded from analysis as follows: stage other than 0–III (69,078); diagnosis by death certificate or autopsy (80) or not microscopically confirmed (369); ineligible histologic type (8166); tumor size unknown, microscopic, diffuse, Paget’s or mammographic report only (42,118); surgery other than unilateral NSM or unilateral non-NSM (262,789); subsequent breast tumor within 2 months of diagnosis (6174); bilateral synchronous breast cancer (20); invalid follow-up (37); or unknown cause of death (1470). After exclusions, 157,592 women were available for analysis, of whom 156,599 (99.4 %) underwent unilateral non-NSM and 993 (0.6 %) unilateral NSM. NSM use increased over time (1988, 0.2 %; 2013, 5.1 %) and with neighborhood SES, and decreased with age (Table 1). The median follow-up was 7.9 years (interquartile range, 3.6–14.0 years) for all patients and for those who had non-NSM, compared to 1.9 years (interquartile range, 0.7–5.5 years) for patients who had NSM (Supplemental Table).
Table 1.
Breast cancer patient characteristics by use of nipple-sparing and non-nipple-sparing unilateral mastectomy in California, 1998–2013
| Variable | Unilateral mastectomy, non-nipple-sparing
|
Unilateral mastectomy, nipple-sparing
|
Total | ||
|---|---|---|---|---|---|
| N | Column % | N | Column % | ||
| All patients | 156,599 | 993 | 157,592 | ||
| Race/ethnicity | |||||
| Non-hispanic white | 103,002 | 65.8 | 634 | 63.8 | 103,636 |
| Non-hispanic black | 8690 | 5.5 | 42 | 4.2 | 8732 |
| Hispanic | 24,368 | 15.6 | 140 | 14.1 | 24,508 |
| Chinese | 4423 | 2.8 | 19 | 1.9 | 4442 |
| Japanese | 2015 | 1.3 | 9 | 0.9 | 2024 |
| Filipina | 6202 | 4.0 | 22 | 2.2 | 6224 |
| Other Asian/Pacific Islander | 6645 | 4.2 | 111 | 11.2 | 6756 |
| Non-hispanic American Indian/other/unknown | 1254 | 0.8 | 16 | 1.6 | 1270 |
| Age at diagnosis, years | |||||
| <40 | 10,537 | 6.7 | 128 | 12.9 | 10,665 |
| 40–49 | 30,174 | 19.3 | 327 | 32.9 | 30,501 |
| 50–64 | 52,090 | 33.3 | 382 | 38.5 | 52,472 |
| 65+ | 63,798 | 40.7 | 156 | 15.7 | 63,954 |
| Marital status at diagnosis | |||||
| Not married | 64,741 | 41.3 | 340 | 34.2 | 65,081 |
| Married | 88,040 | 56.2 | 626 | 63.0 | 88,666 |
| Unknown | 3818 | 2.4 | 27 | 2.7 | 3845 |
| Neighborhood SES statewide quintilea | |||||
| Quintile (Q) 1 (lowest) | 21,610 | 13.8 | 74 | 7.5 | 21,684 |
| Q2 | 29,697 | 19.0 | 127 | 12.8 | 29,824 |
| Q3 | 33,092 | 21.1 | 156 | 15.7 | 33,248 |
| Q4 | 35,417 | 22.6 | 235 | 23.7 | 35,652 |
| Q5 (highest) | 36,783 | 23.5 | 401 | 40.4 | 37,184 |
| Insurance status | |||||
| None | 1330 | 0.8 | 5 | 0.5 | 1335 |
| Private only | 65,849 | 42.0 | 621 | 62.5 | 66,470 |
| Medicare only/Medicare and private | 14,325 | 9.1 | 44 | 4.4 | 14,369 |
| Any public/Medicaid/military | 30,338 | 19.4 | 174 | 17.5 | 30,512 |
| Unknown | 44,757 | 28.6 | 149 | 15.0 | 44,906 |
| American joint committee on cancer stage | |||||
| 0 | 11,791 | 7.5 | 216 | 21.8 | 12,007 |
| I | 50,767 | 32.4 | 374 | 37.7 | 51,141 |
| II | 73,696 | 47.1 | 336 | 33.8 | 74,032 |
| III | 20,345 | 13.0 | 67 | 6.7 | 20,412 |
| Tumor size, cm | |||||
| <1 | 18,206 | 11.6 | 197 | 19.8 | 18,403 |
| 1.0–1.9 | 46,494 | 29.7 | 311 | 31.3 | 46,805 |
| 2.0–2.9 | 37,838 | 24.2 | 211 | 21.2 | 38,049 |
| 3.0–4.9 | 33,361 | 21.3 | 170 | 17.1 | 33,531 |
| 5.0+ | 20,700 | 13.2 | 104 | 10.5 | 20,804 |
| Grade | |||||
| I | 19,675 | 12.6 | 149 | 15.0 | 19,824 |
| II | 55,522 | 35.5 | 396 | 39.9 | 55,918 |
| III | 56,995 | 36.4 | 345 | 34.7 | 57,340 |
| Unknown | 24,407 | 15.6 | 103 | 10.4 | 24,510 |
| Histology | |||||
| Ductal | 132,954 | 84.9 | 845 | 85.1 | 133,799 |
| Lobular or with lobular component | 14,125 | 9.0 | 92 | 9.3 | 14,217 |
| Other | 9520 | 6.1 | 56 | 5.6 | 9576 |
| ER/PR status | |||||
| Negative (ER and PR both negative) | 23,927 | 15.3 | 126 | 12.7 | 24,053 |
| Positive (ER and/or PR-positive) | 92,327 | 59.0 | 715 | 72.0 | 93,042 |
| Unknown/borderline | 40,345 | 25.8 | 152 | 15.3 | 40,497 |
| Lymph node involvement | |||||
| Negative | 91,755 | 58.6 | 735 | 74.0 | 92,490 |
| Positive | 63,369 | 40.5 | 238 | 24.0 | 63,607 |
| Unknown | 1475 | 0.9 | 20 | 2.0 | 1495 |
| Received care at NCI-designated cancer center | |||||
| No | 150,310 | 96.0 | 906 | 91.2 | 151,216 |
| Yes | 6289 | 4.0 | 87 | 8.8 | 6376 |
| Patient SES quintile distributiona of reporting hospital | |||||
| >=50 % of patients in quintiles 4 or 5 (highest) and <50 % in quintiles 1 or 2 | 70,021 | 44.7 | 672 | 67.7 | 70,693 |
| >=50 % of patients in quintiles 1 (lowest) or 2 and <50 % in quintiles 4 or 5 | 35,404 | 22.6 | 129 | 13.0 | 35,533 |
| Mixed SES distribution | 51,174 | 32.7 | 192 | 19.3 | 51,366 |
| Received adjuvant treatment (chemotherapy and/or radiation) | |||||
| No | 91,774 | 58.6 | 547 | 55.1 | 92,321 |
| Yes | 64,825 | 41.4 | 446 | 44.9 | 65,271 |
| Vital status at the end of the study period | |||||
| Alive | 93,815 | 59.9 | 875 | 88.1 | 94,690 |
| Died of breast cancer | 25,948 | 16.6 | 37 | 3.7 | 25,985 |
| Died of another cause | 36,836 | 23.5 | 81 | 8.2 | 36,917 |
|
| |||||
| Variable | Unilateral mastectomy, non-nipple-sparing
|
Unilateral mastectomy, nipple-sparing
|
Total | ||
| N | Row % | N | Row % | ||
|
| |||||
| Year of diagnosis | |||||
| 1988 | 5924 | 99.8 | 14 | 0.2 | 5938 |
| 2013 | 5114 | 94.9 | 276 | 5.1 | 5390 |
ER, estrogen receptor; NCI, National Cancer Institute; PR, progesterone receptor; SES, socioeconomic status
Distribution based on statewide quintiles
In both minimally and fully adjusted models, NSM was associated with lower breast cancer-specific mortality than non-NSM (hazard ratio, HR 0.71, 95 % confidence interval, CI 0.51–0.98 fully adjusted, Table 2). In a secondary analysis limited to diagnoses in 1996 or later, a decreased risk with NSM was seen in the minimally adjusted model (HR 0.61, 95 % CI, 0.38–0.98), but the effect was attenuated in the fully adjusted model (HR 0.79, 95 % CI, 0.48–1.30, data not shown), and further attenuated after adjusting for grade, ER/PR status, and insurance (HR 0.86, 95 % CI, 0.52–1.42).
Table 2.
Breast cancer-specific and overall mortality among patients undergoing nipple-sparing and non-nipple-sparing unilateral mastectomy in California, 1988–2013
| Number of deaths | Total person-years | Age- and stage-adjusteda
|
Fully adjustedb,c
|
|||||
|---|---|---|---|---|---|---|---|---|
| HR | 95 % CI | p value | HR | 95 % CI | p value | |||
| Breast cancer-specific mortality | ||||||||
| 1988–2013 diagnoses | ||||||||
| Non-nipple-sparing | 25,948 | 1451,617 | 1.0a | 1.0b | ||||
| Nipple-sparing | 37 | 4553 | 0.67 | 0.49–0.93 | 0.02 | 0.71 | 0.51–0.98 | 0.04 |
| 1996–2013 diagnoses | ||||||||
| Non-nipple-sparing | 13,469 | 767,098 | 1.0a | 1.0c | ||||
| Nipple-sparing | 17 | 2518 | 0.61 | 0.38–0.98 | 0.04 | 0.86 | 0.52–1.42 | 0.55 |
| Overall mortality | ||||||||
| 1988–2013 diagnoses | ||||||||
| Non-nipple-sparing | 62,784 | 1451,617 | 1.0a | 1.0b | ||||
| Nipple-sparing | 118 | 4553 | 0.91 | 0.76–1.09 | 0.31 | 0.92 | 0.76–1.12 | 0.41 |
| 1996–2013 diagnoses | ||||||||
| Non-nipple-sparing | 29,707 | 767,098 | 1.0a | 1.0c | ||||
| Nipple-sparing | 32 | 2518 | 0.59 | 0.42–0.83 | 0.003 | 0.74 | 0.50–1.08 | 0.12 |
n = 157,592 for 1988–2013; n = 106,181 for 1996–2013
Cox regression with time from diagnosis (days) as the time-scale; stratified by American Joint Committee on Cancer (AJCC) stage (0, I, II, III); and adjusted for age at diagnosis
Cox regression with time from diagnosis (days) as the time-scale; stratified by AJCC stage (0, I, II, III) and histology (ductal, lobular or with lobular component, other); adjusted for age, race, tumor size, lymph node involvement, adjuvant treatment, neighborhood socioeconomic status (SES), marital status, patient SES distribution of reporting hospital, National Cancer Institute-designated cancer center, and year of diagnosis; and adjusted for clustering by hospital
Same as the model in footnote b, but additionally adjusted for grade, estrogen and progesterone receptor status, and insurance status, which were not available before 1996
In both minimally and fully adjusted models, NSM was not associated with overall mortality (Table 2). In a subset with diagnoses in 1996 or later, NSM was associated with lower overall mortality compared with non-NSM in a minimally adjusted model, but the effect was no longer significant after adjustment for all covariates.
Discussion
To the best of our knowledge, this is the largest population-based study of mortality among breast cancer patients treated with NSM compared to non-NSM, with longer median follow-up (7.9 years) than previously reported. Consistent with prior studies [6, 9–14], we found no evidence of worse survival after NSM in this “real world” setting. In fact, NSM was associated with better survival than non-NSM; however, this association did not persist in a multivariable model adjusting for all clinical and sociodemographic factors, including grade, ER/PR status, and insurance status. NSM use increased over time, and was more prevalent among younger women who had earlier-stage cancer and/or resided in higher-SES neighbor-hoods. Thus, the better survival associated with NSM in the minimally adjusted model may reflect confounding by neighborhood SES.
Our study has limitations. Most notably, we had to restrict our assessment to patients having unilateral mastectomy, because SEER and other registries do not capture the nipple-sparing status of bilateral mastectomies. Given the benefits of prophylactic bilateral mastectomy for patients with hereditary breast cancer [5] and the growing interest in bilateral NSM as a less invasive approach for primary breast cancer prevention in high-risk women [13], comparing outcomes of bilateral NSM versus bilateral non-NSM would be clinically valuable. This limitation should be addressed by adding detail about nipple-sparing status to routinely collected registry data items regarding bilateral mastectomy. Other gaps in registry data include family history and inherited genetic mutation status; however, we would not expect major differences in hereditary risk between the two groups that received unilateral mastectomy. Another potential concern is the possibly differential coding of NSM by hospital cancer registrars, which could result in misclassification of some NSM as non-NSM. There was differential follow-up time between patients who received non-NSM compared to NSM; however, the multivariable models that we used controlled for this difference. Moreover, results that included only the more recently diagnosed patients (1996–2013) were similar to those of the full cohort (1988–2013), which offers evidence that our findings are robust to differences in follow-up time. Despite these limitations, however, our study offers considerable strengths: it encompasses the full and diverse population of California, minimizes selection bias and provides results that can be generalized broadly. In the absence of randomized clinical trials, our comprehensive observational study of 157,592 breast cancer patients offers the best available evidence regarding the comparable survival between NSM and non-NSM.
Conclusion
Among California breast cancer patients diagnosed from 1988 to 2013, nipple-sparing mastectomy was not associated with worse survival than non-nipple-sparing mastectomy. These results may inform decisions of patients and doctors deliberating between these surgical approaches for breast cancer treatment.
Supplementary Material
Acknowledgments
This research was funded by the National Cancer Institute’s Surveillance, Epidemiology and End Results Program under contract HHSN261201000140C awarded to the Cancer Prevention Institute of California. The collection of cancer incidence data used in this study was supported by the California Department of Health Services as part of the statewide cancer reporting program mandated by California Health and Safety Code Section 103885; the National Cancer Institute’s Surveillance, Epidemiology, and End Results Program under contract HHSN261201000140C awarded to the Cancer Prevention Institute of California, contract HHSN261201000035C awarded to the University of Southern California, and contract HHSN261201000034C awarded to the Public Health Institute; and the Centers for Disease Control and Prevention’s National Program of Cancer Registries, under agreement #1U58 DP000807-01 awarded to the Public Health Institute. The ideas and opinions expressed herein are those of the authors, and endorsement by the University or State of California, the California Department of Health Services, the National Cancer Institute, or the Centers for Disease Control and Prevention or their contractors and subcontractors neither intended nor should be inferred.
Additional funding sources included the Suzanne Pride Bryan Fund for Breast Cancer Research and the Jan Weimer Junior Faculty Chair in Breast Oncology at Stanford University Cancer Institute.
The funders had no role in the design and conduct of the study; collection, management, analysis, and interpretation of the data; and preparation, review, or approval of the manuscript.
Footnotes
Electronic supplementary material The online version of this article (doi:10.1007/s10549-016-3992-8) contains supplementary material, which is available to authorized users.
Compliance with ethical standards
Conflict of interest The authors declare that they have no conflict of interest.
Ethical standards All research complied with current laws of the United States of America.
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