Parity, hormones and breast cancer subtypes - results from a large nested case-control study in a national screening program

Study population

The Cancer Registry of Norway (CRN) is responsible for the administration of the Norwegian Breast Cancer Screening Program [53]. Women aged 50–69 years are invited to undergo two-view mammography screening every 2 years. From August 2006, women, who underwent mammographic screening in the national program were asked to complete a questionnaire on a number of standard breast cancer risk factors, and another questionnaire on current exposure variables at subsequent screenings. For the current study cohort, women who had participated in the Norwegian Breast Cancer Screening Program during 2006 to 2014 and had completed these questionnaires were eligible for inclusion. The study cohort comprised a total of 344,348 eligible women. Information on cancer cases was obtained through linkage to the population-based CRN records using the unique 11-digit personal identification number assigned to all residents at time of birth. This linkage also included information on vital status, including date and cause of death/date of emigration if applicable. Reporting of cancer to the CRN is mandatory by law, and the registry is considered to be 98.8% complete [54].

All women aged 50–69 years who are included in the National Population Registry are invited to undergo screening as part of the Norwegian Breast Cancer Screening Program every 2 years. The average attendance rate in each round is about 76%.

We conducted a nested case-control study within the study cohort. Only women with no history of ductal carcinoma in situ prior to the study start (1 January 2006) and no history of diagnosis of another invasive cancer (except non-melanoma skin cancer) were eligible for the study. The cases were women diagnosed with a first occurrence of invasive breast cancer (ICD10: C50) during the study period, with information on ER, PR, and HER2 status (see subsequent text). For each woman with breast cancer, we randomly selected five controls individually matched to cases by year of birth (+/- 3 years) and year of last screening before breast cancer diagnosis (+/- 3 years). Controls had to be alive and resident in the country at the time of the diagnosis of breast cancer in the matching case. We ended up with 6471 patients with breast cancer (cases) and 32,355 controls. The study was approved by the Regional Committee for Medical and Health Research Ethics in the South-East Health Region of Norway.

Tumor receptor status ascertainment

Information on ER, PR and HER2 status was assessed by IHC and extracted from pathology reports submitted to the CRN. From 2006 to January 2012, tumors were classified as ER-negative if there was <10% reactivity. From February 2012 onwards, the threshold for ER-negative tumors was changed to <1% reactivity as a result of change in the treatment protocols for patients attending clinics in Norway. We used these official thresholds. PR-negative tumors were defined as those with reactivity of <10%, and PR-positive tumors as those with reactivity ≥10%. HER2 expression status was determined at each laboratory with IHC and/or in situ hybridization. Cases with no (0) or weak (1+) immunostaining were defined as HER2-negative, while cases with moderate (2+) or strong immunostaining (3+) were defined as HER2-positive. In situ hybridization (fluorescence (FISH), chromogenic (CISH), or silver (SISH) in situ hybridization methods) was usually used to confirm HER2 status if IHC yielded 2+ results. If IHC was 2+ and FISH, CISH, or SISH were missing, or if IHC was missing but FISH, CISH, or SISH were positive, the tumor was classified as HER2-positive. If IHC was 2+ and FISH, CISH, and SISH were negative, the tumor was regarded as HER2-negative. Data on Ki-67 were recorded by the CRN from late 2011 and therefore, we did not include this marker in our analysis.

Risk factors

Our primary exposures of interest were hormonal risk factors including reproductive factors (age at first birth, number of pregnancies, breastfeeding, menopausal status), and other hormonal factors (use of oral contraceptives, intrauterine devices (IUD) and menopausal hormone therapy use). Menopausal hormone therapy use included use of estrogen alone (estrogen therapy, ET), or use of combined estrogen and progestin therapy (EPT). For current exposures we chose the questionnaire at the last screening before breast cancer diagnosis for the cases and the corresponding time for the controls. If the questionnaire or any values were missing, information was used from the previous screening questionnaire. Menopausal status was defined according to whether or not women were still menstruating, or whether they menstruated irregularly. Menopausal age was defined as the age when menstruation ended.

Confounders and missing values

We considered age at screening (50–54, 55–59, 60–64, or 65–70 years), body mass index (BMI) at screening (≤22, 23–25, 26-28, or >28), education (no education/primary school, high school, or bachelor’s/master’s degree), age at menarche (9–12, 13, 14, or 15–18 years), number of pregnancies lasting at least six months (never pregnant, 1, 2, 3, or ≥4) and menopausal status (premenopausal, perimenopausal, or postmenopausal) as potential confounders and adjusted for these when appropriate. We tried to control more tightly for age at screening using a continuous variable, but the results remained largely the same, and we therefore retained the 5-year categories.

Women with missing values for an exposure variable were excluded from the analyses of that variable, while women with missing information on the potential confounding variables listed previously were excluded from all analyses. Of the 6471 patients with breast cancer, we excluded women due to missing information on the following variables: BMI (n = 532 patients), age at menarche (n = 298 patients), education (n = 66 patients), number of pregnancies (n = 164 patients), and menopausal status (n = 59 patients). This left us with 5352 women with breast cancer (cases) for analysis. Of the 32,355 controls, we excluded controls based on missing information on the following: BMI (n = 3296 controls), age at menarche (n = 1641 controls), education (n = 371 controls), number of pregnancies (n = 779 controls), and menopausal status (n = 336 controls). This left us with 25,932 controls for analysis.

The cases were categorized by breast cancer subtype using a modified version of the classification of clinically defined subtypes proposed at the St. Gallen meeting in 2013 [46]. Of the 5352 cases of breast cancer, 604 had unknown hormone receptor (i.e. ER and/or PR) and HER2 status or could not be classified into the breast cancer subtypes. There were 4748 women with breast cancer classified into the following subtypes: 2985 women with luminal A-like breast cancer (ER+ PR+ HER2-), 758 women with luminal B-like HER2-negative breast cancer (ER+ PR- HER2-), 396 women with luminal B-like HER2-positive breast cancer (ER+ PR+/PR- HER2+), 223 women with HER2-positive breast cancer (ER- PR- HER2+) and 386 women with triple-negative breast cancer (ER- PR- HER2-). As we did not have Ki-67 results, we conducted sensitivity analysis where we added grade to the luminal subtype definitions in an attempt to better separate out these subtypes, using the definitions from the St. Gallen 2013 meeting [46]. In this analysis luminal A-like subtype was defined as ER+ PR+ HER2-, low or medium grade, luminal B-like HER2-negative as ER+ PR- HER2-, high grade, and luminal B-like HER2-positive as ER+ PR-/PR+ HER2+, any grade.

Statistical analyses

We used conditional logistic regression to estimate odds ratios (with 95% confidence intervals (CI)) as a measure of relative risk associated with various exposures. Trend tests were performed by fitting ordinal values corresponding to exposure categories and testing whether the slope coefficient differed from zero. To test for heterogeneity between breast cancer subtypes we ran case–case analyses, comparing each subtype to the luminal-A-like subtype. We used likelihood ratio tests comparing the likelihood ratio of the case–case model with confounders only to that of the adjusted case–case model with the exposure variable. We considered a two-sided p value <0.05 as statistically significant.

Because we used conditional logistic regression we included all women in the risk estimates (e.g. women who were never pregnant were included in analysis of age at first birth, women who never used oral contraceptives (OC) were included in analysis of age at the start of OC, and premenopausal women were included in analysis of age at menopause), but they were not included in the trend test (dummy variables were added into the analysis).

Reproductive factors

Our observation of an inverse association between parity and luminal A-like breast cancer is consistent with the vast majority of studies as summarized in the recent systematic review of 38 studies [47] and the more recent studies that were not included in the review [7, 8, 10, 48–51]. The results for luminal B breast cancer have been less clear, with studies finding a protective or increased effect, or no effect of parity [47–50]. In the current study, parity was associated with a decreased risk of both luminal B-like subtypes. This is consistent with four out of six studies included in the systematic review [47] and three of the more recent studies [48–50]. Two of these recent studies had information on Ki-67 [48, 49]. When we added grade to the luminal A-like and luminal B-like subtypes, the parity results remained largely the same. These results suggest that there is a protective effect against luminal B, and the effect seems apparent regardless of the markers used to define the luminal B subtype.

Although not statistically significant, we found that parity was associated with a decreased risk of HER2-positive breast cancer. This is consistent with the findings in the Nurses’ Health Study [49], but inconsistent with a case–control study from the Breast Cancer Family Registry [50] and a Korean cohort study [48], which reported that parity was associated with an increased risk of the HER2-positive subtype. One of these latter studies included very few women with HER2-positive breast cancer and both studies included younger women than the current study.

However, although not statistically significant, we found that parity was associated with a decreased risk of triple-negative breast cancer.

This is inconsistent with several other studies in which parity was associated with no risk [48], or was associated with increased risk of triple-negative breast cancer [7, 47, 50]. One of the studies was in African American women who are less likely to breastfeed compared to Caucasian women [7]. Several studies have used CK5/6 and epithelial growth factor receptor (EGFR) in addition to ER, PR and HER2neu to define a basal-like subtype [9, 39, 40, 49, 56], and this may be one reason for the inconsistency between our study and these other studies. We included 386 women with triple-negative breast cancer in our study, but few of these had never been pregnant (n = 34) or had four or more children (n = 36). More notably, our study consisted of older women (aged 50–69 years). Although we did not have information on time since the last pregnancy, we can only speculate that few women in our study had had a recent pregnancy, a factor associated with increased risk of triple-negative cancer [57]. Perhaps this explains why we did not find an increased risk of triple-negative cancer with multi-parity.

We found that age at first birth (26 years or older) was associated with an increased risk of luminal-like breast cancer. This is consistent with a systematic review and [47] and two recent studies [8, 49]. However, age at first birth was only statistically significantly associated with luminal A cancer, whereas in the current study age at first birth was significantly associated with luminal A-like and luminal B-like HER2-negative subtypes. One reason for this difference might be that our definition of luminal B-like HER2-negative cancer was rather similar to the definition of the luminal A-like subtype used in the other studies. In the current study, a non-statistically significant positive association was observed between late age at first birth (31 years or older), and both HER2-positive and triple-negative breast cancer. This is consistent with the Nurses’ Health Study [49], and inconsistent with the case-control study from Korea [8]. In the latter study, the majority of women were premenopausal, whereas the current study included mainly postmenopausal women.

We found no protective effect of breastfeeding on breast cancer occurrence overall or for any subtype. This is inconsistent with the systematic review that reported that breastfeeding is associated with decreased risk of luminal A-like, luminal B-like and triple-negative subtypes [47].This may be because our study only included women above 50 years of age, and is consistent with the suggestion that the protective effect of breastfeeding is relatively time-limited, and may be seen predominantly in younger women [58–60]. Several of the more recent studies found a significant protective effect against basal-like breast cancer [7, 49–51]. These studies included both triple-negative breast cancer as in our study and basal-like breast cancer with more markers (CK 5/6 and EGFR) [49].

Oral contraceptives, intrauterine devices, and menopausal hormone therapy use

For long duration of oral contraceptive use (>10 years), we observed a slight increase in risk of all the subtypes except for the luminal B-like HER2-positive and HER2-positive subtypes. Our positive association between duration of oral contraceptive use and triple-negative breast cancer is consistent with the systematic review [47]. However, two of the three studies in the review reported a decreased risk between oral contraceptive use and the luminal A subtype [9, 34]. These latter studies were smaller and included younger women (20–74 and <56 years old) than the current study. There are few data on IUD use. Two studies reported that IUDs were not associated with an increased risk of breast cancer [61, 62], whereas other studies report that IUDs were associated with an increased breast cancer risk [63, 64]. We observed a significant positive trend for association between duration of IUD use and breast cancer overall and luminal A-like breast cancer. However, the increased risk of luminal A-like breast cancer was only statistically significant in women using IUDs for more than 10 years. This is consistent with two Finnish studies of levonorgestrel-releasing IUDs and breast cancer [63, 64], which reported that levonorgestrel-releasing IUDs were associated with an increased risk of breast cancer. One hypothesis is that levonorgestrel-releasing IUDs have substantial progestogenic and androgenic effects [63, 64], which could contribute to this increased risk with IUD use.

We observed a large increased risk of luminal A-like and a moderate increased risk of luminal B-like breast cancer with use of menopausal hormone therapy, albeit only significantly so for the luminal A-like subtype. This is consistent with a systematic review [47] and the cohort study from the Nurses’ Health Study that investigated only one luminal B subtype, and used Ki-67 to differentiate between luminal A-like and luminal B-like subtypes [49]. There was some evidence that menopausal hormone therapy was associated with a slightly decreased risk of triple-negative breast cancer when we compared current menopausal hormone therapy users to never users. This is consistent with one of the studies in the systematic review, which was a case-control study from Washington state [35]. The latter study used the same definition of triple-negative breast cancer as the current study. Inconsistent with our result, studies from the Women’s Health Initiative [65] and the Nurses’ Health Study [40, 49] reported an association between menopausal hormone therapy and increased risk of triple-negative breast cancer. In these latter studies, they used more biomarkers (CK 5/6 and EGFR) to define the basal-like subtype and included younger women than the current study. When we looked at the duration of EPT, compared to never users, women who had used EPT for more than 5 years were at an increased risk of HER2-positive and triple-negative breast cancer. This is consistent with the analysis from the Nurses’ Health Study [49].

We hypothesized that EPT would have a stronger effect on thin women than on heavier women, and our results suggested modification of the effect of the duration of EPT when we analyzed leaner women (BMI <25) and heavier women (BMI >25) with the luminal-like subtypes of breast cancer. This is consistent with a previous population-based case–control study of women aged 55–72 years [66].

Mechanisms and suggested subtype differences

Pregnancies have been reported to ultimately reduce plasma estrogen (estrone, estradiol and estriol) [67] and follicular-phase progesterone [68], and increase sex hormone-binding globulin [67]. Our findings suggest that both parity and combined menopausal hormone therapy may be predominantly associated with luminal-like breast tumors, with the association being strongest for the luminal A-like subtype. The effect of EPT was stronger when we added grade to the definition of luminal A cancer. We observed little effect of breastfeeding overall, and no clear subtype differences. It is possible that the effect of breastfeeding is non-hormonal, and includes changes in immune responses and apoptosis [69, 70].

Our results, together with those from other large studies, further suggest that the associations between these hormonal-related factors (parity, age at first birth, oral contraceptive use and menopausal hormone therapy use) and risk of the luminal B-like subtypes are similar to the associations between these factors and risk of the luminal A-like subtype. However, although the association with luminal A was the strongest, it is clear that EPT also increases the risk of breast cancer with bad prognosis [71].

We observed some intriguing associations with HER2-positive breast cancer. The effect of pregnancies and age at first birth were similar for the HER2-positive subtype and luminal cancer. Also, age at menopause was strongly associated with increased risk only of the HER2-positive subtype. Further, we observed a positive trend of association between age at menopause and the luminal B-like HER2-positive subtype. These results might imply that there are hormonal mechanisms involved in the expression of the HER2 protein. On the other hand, exogenous hormone use (EPT) was not associated with this subtype, suggesting that perhaps only some hormonal mechanisms play a role in HER2-positive subtypes. The contrasting results between the luminal-like subtypes and triple-negative breast cancer are also consistent with previous literature. Hormonal factors have a stronger effect on ER+ PR+ tumors, which suggests that the etiology of triple-negative cancer is different from that of the luminal subtypes. Specifically, this suggests that triple-negative tumors may not be as easily prevented hormonally.

In summary, the strongest discrepancy across subtypes was for the use of combined hormone therapy, where the effect was clearly much stronger for luminal A-like than for other cancers. Otherwise, the associations with hormonal risk factors were stronger for luminal A and B-like subtypes than for HER2-positive and triple-negative subtypes. Our results suggest that reproductive factors may to some extent be associated with HER2-positive tumors, but that triple-negative tumors have a different etiology.

Strengths and limitations

Strengths of this study include its population-based design, the large size, being the largest single study of breast cancer subtypes conducted so far, and the availability of prospectively collected detailed information on many risk factors for breast cancer.

Although the study is the largest to date, there was still limited power for analysis of the rare breast cancer subtypes. Another weakness was the lack of molecular expression data. This may have obscured differences between the subtypes. Another possible weakness was that a subset of women were diagnosed within a month of completing the questionnaire (n = 1813), and could have been symptomatic when they completed their questionnaire. However, exclusion of women who were diagnosed within a month of completing the questionnaire did not affect the results (results not shown).

Women who attend screening might be more health-conscious and have a healthier lifestyle than women who do not attend. This could have contributed to obliteration of protective effects of “healthy” habits, such as an effect of breastfeeding overall. At the same time, women who attend screening are more likely to have breast cancer detected. Thus, the picture becomes complicated with these potential biases, and it is not clear how this would explain the results of this paper. The associations between well-established risk factors and overall risk of breast cancer were largely as expected. Furthermore, it is unlikely that any such bias would have differentially affected the results for different subtypes.