Effect of hysterectomy on risk of thyroid cancer and benign thyroid disorders in South Korean women.

We included 13,498 participants who fulfilled our criteria in the groups with and without hysterectomy (Fig.  1 ). They were 47 [45–50] years old, had a follow-up of 11.4 [10–13.5] years, and a BMI of 23.6 [21.8–25.7] kg/m 2 . Table  1 shows the detailed characteristics of these participants. Supplementary Table 1 shows the detailed characteristics of the participants before propensity score matching. After propensity score matching, the baseline prevalence of major estrogen-dependent diseases, namely endometriosis and uterine fibroids, was well-balanced between the hysterectomy and control groups (endometriosis: 17.3% vs. 17.3%; uterine fibroids: 73.8% vs. 73.5%, respectively; Table  1 ). Accordingly, the subsequent risk analyses for thyroid cancer and benign thyroid disorders were adjusted for these baseline variables. Fig. 1 Flowchart showing the process of selecting hysterectomy and non-hysterectomy groups using 2002–2020 Korean National Health Insurance data. Flowchart showing the process of selecting hysterectomy and non-hysterectomy groups using 2002–2020 Korean National Health Insurance data. Table 1 Characteristics of hysterectomized and non-hysterectomized women selected after propensity score matching in our study. Non-Hysterectomy Hysterectomy Total P -value Standardized mean difference Number of women 13,498 13,498 26,996 Follow-up period (years) 11.4 [10-13.6] 11.5 [10-13.4] 11.4 [10-13.5] 0.869 0.026 Median age (years) 48 [45–50] 47 [45–50] 47 [45–50] 0.461 0.014 Age at inclusion (years) < 0.001 0.076 40 –44 3,369 (25) 3,193 (23.7) 6,562 (24.3) 45–49 5,953 (44.1) 6,436 (47.7) 12,389 (45.9) 50–55 3,421 (25.3) 3,229 (23.9) 6,650 (24.6) 55–60 755 (5.6) 640 (4.7) 1,395 (5.2) Year at inclusion 0.371 0.006 2003–2005 1,585 (11.7) 1,584 (11.7) 3,169 (11.7) 2006–2008 4,607 (34.1) 4,645 (34.4) 9,252 (34.3) 2009–2011 7,306 (54.1) 7,269 (53.9) 14,575 (54) Median BMI (kg/m 2 ) 23.6 [21.7–25.7] 23.6 [21.8–25.7] 23.6 [21.8–25.7] 0.388 0.004 BMI (kg/m 2 ) 0.059 0.017 < 18.5 199 (1.5) 199 (1.5) 398 (1.5) 18.5–22.9 5,357 (39.7) 5,352 (39.7) 10,709 (39.7) 23–24.9 3,437 (25.5) 3,502 (25.9) 6,939 (25.7) 25–29.9 3,918 (29) 3,897 (28.9) 7,815 (28.9) ≥ 30 587 (4.3) 548 (4.1) 1,135 (4.2) Low SES 49 (0.4) 61 (0.5) 110 (0.4) 0.195 0.014 Rural area 9,664 (71.6) 9,675 (71.7) 19,339 (71.6) 0.892 0.002 CCI 0.33 0.009 0 11,032 (81.7) 11,063 (82) 22,095 (81.8) 1 2,028 (15) 1,988 (14.7) 4,016 (14.9) ≥ 2 438 (3.2) 447 (3.3) 885 (3.3) Parity < 0.001 0.033 0 or not respond 1,988 (14.7) 2,126 (15.8) 4,114 (15.2) 1 1,710 (12.7) 1,683 (12.5) 3,393 (12.6) 2 8,991 (66.6) 8,938 (66.2) 17,929 (66.4) ≥ 3 809 (6) 751 (5.6) 1,560 (5.8) Age over 13 at menarche (years) 10,528 (78) 10,473 (77.6) 21,001 (77.8) 0.067 0.010 Menopause before inclusion 2,068 (15.3) 2,025 (15) 4,093 (15.2) 0.023 0.009 Smoking 0.094 0.017 Never 12,776 (94.7) 12,741 (94.4) 25,517 (94.5) Past 206 (1.5) 198 (1.5) 404 (1.5) Current 516 (3.8) 559 (4.1) 1,075 (4) Alcohol (per week) < 0.001 0.068 None 9,623 (71.3) 9,491 (70.3) 19,114 (70.8) ~ 2/week 3,683 (27.3) 3,693 (27.4) 7,376 (27.3) 3–6/week 138 (1) 234 (1.7) 372 (1.4) Daily 54 (0.4) 80 (0.6) 134 (0.5) Physical exercise (per week) 0.386 0.025 None 8,895 (65.9) 8,780 (65) 17,675 (65.5) 1–2 2,468 (18.3) 2,548 (18.9) 5,016 (18.6) 3–4 1,309 (9.7) 1,315 (9.7) 2,624 (9.7) 5–6 398 (2.9) 386 (2.9) 784 (2.9) Daily 428 (3.2) 469 (3.5) 897 (3.3) DM 1,229 (9.1) 1,263 (9.4) 2,492 (9.2) 0.477 0.009 Hypertension 2,538 (18.8) 2,501 (18.5) 5,039 (18.7) 0.209 0.007 Dyslipidemia 1,906 (14.1) 1,838 (13.6) 3,744 (13.9) 0.218 0.015 SLE 11 (0.1) 11 (0.1) 22 (0.1) 1 < 0.001 IBD 3,099 (23) 2,961 (21.9) 6,060 (22.4) 0.046 0.025 Crohn’s disease 9 (0.1) 4 (0) 13 (0) 0.267 0.017 MHT 227 (1.7) 163 (1.2) 390 (1.4) < 0.001 0.040 Adnexal surgery 150 (1.1) 123 (0.9) 273 (1) 0.019 0.020 Uterine fibroids 9,962 (73.8) 9,922 (73.5) 19,884 (73.7) < 0.001 0.007 Endometriosis 2,337 (17.3) 2,336 (17.3) 4,673 (17.3) 1 < 0.001 BMI, body mass index; DM, diabetes mellitus; CCI, Charlson comorbidity index; IBD, inflammatory bowel disease; MHT, menopausal hormone therapy; SES, socioeconomic status; SLE, systemic lupus erythematosus The data is shown as a number (%) or median value (interquantile range). Characteristics of hysterectomized and non-hysterectomized women selected after propensity score matching in our study. BMI, body mass index; DM, diabetes mellitus; CCI, Charlson comorbidity index; IBD, inflammatory bowel disease; MHT, menopausal hormone therapy; SES, socioeconomic status; SLE, systemic lupus erythematosus The data is shown as a number (%) or median value (interquantile range). During a median follow-up of 11.4 years, thyroid cancer was newly diagnosed in 3.4% of the hysterectomy group compared to 2.6% of the non-hysterectomy group ( p  < 0.001) (Fig.  2 ). For benign thyroid disease, incidence was 11.9% in the hysterectomy group versus 11.1% in controls (p-value 0.054). The incidence of thyroid cancer per 100,000 person-years is presented in Supplementary Table 2. Fig. 2 Risk analysis of hysterectomy for thyroid cancer and individual thyroid diseases in our study. CI, confidence interval; HR, hazard ratio. Risk analysis of hysterectomy for thyroid cancer and individual thyroid diseases in our study. CI, confidence interval; HR, hazard ratio. Moreover, the survival analysis of thyroid cancer incidence between the two groups indicated a statistically significant increase in the hysterectomy group compared to the non-hysterectomy group (stratified log-rank test: p -value < 0.001). The Kaplan-Meier plot illustrating the survival analysis is depicted in Fig.  3 . Fig. 3 Kaplan-Meier plots of thyroid cancer incidence in the hysterectomy and non-hysterectomy groups (stratified log-rank test p -value < 0.001). Kaplan-Meier plots of thyroid cancer incidence in the hysterectomy and non-hysterectomy groups (stratified log-rank test p -value < 0.001). In our stratified-extended Cox proportional analysis, the risk of thyroid cancer was found to be higher in the hysterectomy group compared to the non-hysterectomy group (HR 1.351, 95% CI 1.17–1.559, p  < 0.001) (Fig.  2 ). Specifically, hysterectomy without adnexal surgery was associated with an increased risk of thyroid cancer (HR 1.361, 95% CI 1.165–1.591, p  < 0.001), while the risk of thyroid cancer in hysterectomy with adnexal surgery did not differ significantly from the non-hysterectomy group (HR 1.291, 95% CI 0.898–1.865, p  = 0.117). Regarding overall benign thyroid disease, the risk was higher in the hysterectomy group compared to the non-hysterectomy group within ten years after the procedure (HR 1.086, 95% CI 1.031–1.145, p  = 0.002). However, the risk did not differ significantly beyond ten years after hysterectomy (HR 0.775, 95% CI 0.593–1.014, p  = 0.063). Furthermore, the risk of goiter (HR 1.311, 95% CI 1.099–1.562, p  = 0.004) and nontoxic single thyroid nodule (HR 1.134, 95% CI 1.007–1.278, p  = 0.038) was higher in the hysterectomy group compared to the non-hysterectomy group. However, no significant difference was observed for other benign thyroid conditions between the two groups. Supplementary Table 3 also shows the risk of thyroid cancer and individual thyroid disease by age at hysterectomy. Age-stratified risk analysis demonstrated that hysterectomy was associated with an increased risk of thyroid cancer particularly in women aged 40–49 and 50–59 years. Specifically, for thyroid cancer, the HR was 1.336 (95% CI, 1.123–1.590; P  = 0.001) in the 40–49 years group and 1.423 (95% CI, 1.065–1.901; P  = 0.017) in the 50–59 years group, indicating significantly higher risk compared to non-hysterectomized women of the same age. For benign thyroid diseases, the association varied by subtype and age group. In women aged 50–59 years, hysterectomy was significantly associated with increased risks of goiter (HR 1.459; 95% CI, 1.053–2.021; P  = 0.023), nontoxic single thyroid nodule (HR 1.277; 95% CI, 1.015–1.606; P  = 0.037), and total benign thyroid disease (HR 1.237; 95% CI, 1.076–1.423; P  = 0.003). In contrast, these associations were not statistically significant in the 40–49 years group. No significant age-stratified associations were observed for hypothyroidism, hyperthyroidism, autoimmune thyroid disease, or other thyroid diseases in either age group. In a sensitivity test including only women with a CCI of 0, hysterectomy was associated with a similar risk of thyroid cancer (HR 1.417, 95% CI 1.204–1.668) and benign thyroid disease (< 10 years: HR 1.073, 95% CI 1.01–1.141) (≥ 10 years: HR 0.673, 95% CI 0.494–0.917) to the main outcome.

This study analyzed data from the NHIS, which provides health insurance to most of the South Korean population (97%) 16 . The NHIS data contain information on sociodemographic factors, medical diagnoses based on the International Statistical Classification of Diseases, Tenth Revision (ICD-10) codes, and treatment data for the South Korean population 16 . The NHIS also performs biennial health checkups for all employees or citizens aged 40 and older in South Korea 16 . The data include laboratory tests, self-reported lifestyle habits, and anthropometric measurements from these checkups 16 . The study used a retrospective cohort design and the data spans from 2002 to 2020. This medical study enrolled women aged 40–59 years who underwent hysterectomy due to benign conditions between January 1, 2003, and December 1, 2011, comprising the hysterectomy group. The non-hysterectomy group consisted of women within the same age range who had a health checkup at the NHIS during the corresponding period. To accommodate the capacity of the NHIS analytic server, we randomly selected 25% of the eligible women. To ensure a clean study population, we conducted a washout process, excluding women who had undergone a health checkup or hysterectomy in 2002. Additionally, participants who received a diagnosis code for cancer (any Cxx) from a medical institution within 365 days of joining the study and those who received a diagnosis code for any thyroid disease (E00–E07) within 180 days of joining the study were also excluded. The comparison groups were established using 1:1 propensity score matching, taking into account various important variables, such as age, smoking status, physical exercise level, alcohol consumption, dyslipidemia, diabetes mellitus, hypertension, socioeconomic status (SES), body mass index (BMI), age at menarche, age at menopause, parity, residential area, Charlson comorbidity index (CCI) score, systemic lupus erythematosus (SLE), inflammatory bowel disease (IBD), Crohn’s disease, prior history of menopausal hormone therapy (MHT), uterine fibroids, previous adnexal surgery, and endometriosis. Throughout the study duration, which continued until December 31, 2020, the participants were closely monitored and observed. Thyroid cancer was diagnosed if there were at least three visits to a medical institution with a related diagnosis code (C73). The study identified benign thyroid disease as having both thyroid-related tests (thyroid scan, thyroid function test) and thyroid disease diagnosis codes (E03 –E07). The study excluded benign thyroid disease if the number of visits to a medical institution for thyroid disease was less than three. The classification of thyroid disorders is based on the level of thyroid hormone production, the size and shape of the gland, the presence of autoimmunity, the presence of nodules, and the malignancy. The categories are: hyperthyroidism disorders (E05.0, E05.1, E05.2, E05.5, E05.8, E05.9), hypothyroidism disorders (E03.4, E03.5, E03.8, E03.9), autoimmune thyroid disorders (E06.2, E06.3, E06.5, E06.9), enlarged thyroid gland disorders (E04.0, E04.2, E04.8, E04.9), nontoxic single thyroid nodule disorders (E04.1), other thyroid disorders (E07.8, E07.9), and thyroid cancer (C73). CCI score was calculated based on diagnosis codes from one year before the study participation date until the participation date 17 . Adnexal surgeries were recorded using specific surgery codes. Also, this study investigated various factors, including age (grouped in 5-year increments), self-reported measures of smoking, alcohol drinking, and physical activity levels, SES (referred to as medical aid for medical insurance), and residential area (rural or urban area). Parity was grouped into four categories: 0, 1, 2, and ≥ 3 births. BMI was calculated using the Asia-Pacific perspective criteria 18 . Menarchial age was divided into two groups: <13 years and ≥ 13 years. Menopausal status was determined through questionnaire responses. MHT was defined as the use of such therapy (estrogen/progestogen, estrogen, and tibolone) for more than six months before joining the study. The presence of hypertension (I10–I15), diabetes mellitus (E10–E14), hyperlipidemia (E78), uterine fibroids (D25), and endometriosis (N80) was ascertained based on whether the subjects had visited a medical institution for the respective conditions two or more times before study participation. For SLE (M32), IBD (K58), and Crohn’s disease (K50), a diagnosis code being recorded three or more times in medical visits defined the presence of these conditions. Baseline characteristics were compared before propensity score matching using appropriate statistical tests: Student’s t-test (or Wilcoxon rank-sum test) for continuous variables and χ 2 test (or Fisher’s exact test) for categorical variables. After propensity score matching, paired t-test (or Wilcoxon signed-rank test) was employed to compare continuous variables, while the Cochran-Mantel-Haenszel test was used for categorical variables. Data were presented as median (interquartile range) for continuous variables and as value (percentage) for categorical variables. The incidence of thyroid disease was determined as the number of new cases per 100,000 person-years during the follow-up period. To assess the probability of being free of thyroid cancer, Kaplan-Meier curves were utilized, and group differences were analyzed using the stratified log-rank test. For estimating hazard ratios (HRs) and 95% confidence intervals (CIs) regarding thyroid disease incidence, we employed Cox proportional hazards regression models. The proportional hazards assumption was assessed using the Schoenfeld residual test. In cases where the assumption was not met, an extended Cox analysis was conducted using a step function. The study entry date for the hysterectomized group was based on the day of the hysterectomy, while for the non-hysterectomized group, it was determined as the first day of health checkup. The censoring date was defined as the earliest occurrence of death or the date of the last NHIS-recorded healthcare encounter (outpatient, inpatient, or routine screening/health checkup), not limited to illness visits. Missing values were removed through the list deletion method during propensity score matching. Furthermore, exploratory subgroup analyses were performed based on age. To verify the robustness of the original results, additional Cox proportional hazards regression analyses were conducted on women with a CCI of zero. All statistical analyses were conducted using R statistical software version 3.5.1 (The R Foundation for Statistical Computing, Vienna, Austria). Statistical significance was set at a two-sided P-value < 0.05. The normality of continuous variables was assessed by balancing individual covariates matched by standardized mean differences and the Anderson-Darling normality test. For interpretation, standardized mean differences (SMD) in Supplementary Table 1 (pre-matching) use a threshold of < 0.1 to indicate good covariate balance; values above this highlight imbalances addressed by propensity score matching. Data analysis was carried out between July 2022 and January 2023. This study was approved by the Institutional Review Board of Inje University Sanggye Paik Hospital (approval number: SGPAIK 2021-12-005). As these data are public and nonpersonally identifiable, the Institutional Review Board of Inje University Sanggye Paik Hospital IRB waived informed consent. The NHIS carefully anonymized the raw data and performed data analysis within its secure closed server, following its personal protection policy. To maintain confidentiality, unauthorized access to raw data was prevented by restricting data export solely to research outcome analysis. This measure ensured that only authorized individuals could access data and maintained confidentiality and integrity. Informed consent was not required for this study, in accordance with the Bioethics and Safety Act of South Korea. All methods were performed in accordance with the relevant guidelines and regulations.

In this large, population-based cohort study of South Korean women, we found that hysterectomy was associated with a significantly increased risk of thyroid cancer (HR 1.35, 95% CI 1.17–1.56, p  < 0.001) and certain benign thyroid disorders, particularly goiter and nontoxic single thyroid nodule, after propensity score matching. The heightened risk of thyroid cancer was most notable in women aged 40–59, and persisted regardless of whether adnexal surgery was performed. No significant associations were found between hysterectomy and risk of hypothyroidism, hyperthyroidism, or autoimmune thyroid disease. According to previously mentioned epidemiological data and experimental studies, we considered estrogen a potent growth factor for benign and malignant thyroid cells, which can explain the sex difference in the prevalence of thyroid nodules and thyroid cancer. We expected that early menopause after a simple hysterectomy or hysterectomy with BSO has a protective effect on thyroid disease development, including thyroid cancer. However, population-based studies, including our study, reported a significantly higher incidence of thyroid cancer after hysterectomy without adnexectomy. Additionally, the risk of goiter was elevated with hysterectomy without adnexal surgery, and hysterectomy with adnexal surgery was linked to a higher risk of non-toxic single thyroid nodules. These findings may be interpreted that an abrupt or early gradual decline in estrogen levels is not a protective factor in the development of benign thyroid disease and cancer. While estrogen has long been considered a common factor affecting both uterine and thyroid conditions, recent studies call for broader consideration. Machine learning-based research has identified links among endometriosis, benign breast disease, and non-toxic goiter, suggesting these ties may be more related to iodine than to estrogen alone 19 . Iodine deficiency stimulates increased secretion of thyroid-stimulating hormone (TSH) 20 , which is a key driver of thyroid cell growth 21 . Areas with low iodine intake tend to have higher rates of goiter, thyroid nodules, and papillary thyroid cancer 22 , 23 . Unlike goiter, iodine deficiency promotes estrogen-induced transcription by reducing the activity of BRCA1, an inhibitor of ERα transcription, and reduces the metabolism of estrone or estradiol by decreasing cytochrome P450 1A1 and 1B1 24 . This “estrogenic up-regulation” is thought to promote benign breast conditions and may also influence estrogen-dependent uterine diseases, such as uterine fibroids, adenomyosis, endometrial hyperplasia, and endometrial cancer. However, limited research has been conducted on the association between uterine disease and iodine deficiency. There is emerging evidence linking uterine fibroids (a common reason for hysterectomy) with thyroid nodules 25 , and abnormal uterine bleeding with thyroid dysfunction 26 . This suggests that iodine deficiency could be a common thread increasing disease risk in the thyroid, breast, and uterus. The co-occurrence of these conditions may also reflect a tendency for individuals with one iodine deficiency–related disease to be diagnosed or treated for others, but more research is needed to clarify these relationships. Our findings are consistent with several prior studies reporting an increased risk of thyroid cancer following hysterectomy, both with and without oophorectomy 27 – 29 . Some population-based studies have noted a similar magnitude of risk elevation, although results have varied according to ethnic background, confounder adjustment, and study design 12 , 13 , 30 , 31 . A meta-analysis of three observational cohort studies found a summary relative risk of 1.56 (95%CI: 1.30–1.87) for the association between hysterectomy and thyroid cancer 32 . Few studies have examined the association between hysterectomy and benign thyroid disorders; our data suggest a modestly increased risk, particularly for goiter and thyroid nodules. In Supplementary Table 3, some findings showed borderline significance, such as the risk of goiter in women aged 40–49 after hysterectomy (HR 1.229, 95% CI 0.987–1.532, P = 0.066). This suggests a possible increased risk, but the association was not statistically significant. These borderline results may be due to small sample sizes, modest effect sizes, or residual confounding, and should be interpreted with caution. Additional research is needed to clarify these trends. This study has several strengths. Most studies have been conducted in Western countries, including the USA, Sweden, Finland, Australia, and France. In contrast, our study pioneered the study in Asian populations. South Korea is regarded as a region with iodine sufficiency based on national nutrition surveys, indicating that iodine deficiency is unlikely a major confounding factor in this cohort. Notably, South Korea’s NHIS has established a well-structured data computerization process for research purposes, enabling the use of high-quality big data, including a substantial cohort with a long follow-up period. We excluded women with thyroid disease during their initial hysterectomy to minimize the potential impact of confounding factors. Furthermore, we employed propensity score matching to control for confounding biases by adjusting for various traditional medical histories, such as age, smoking status, physical exercise level, alcohol consumption, SES, BMI, age at menarche, age at menopause, parity, residential area, dyslipidemia, diabetes mellitus, hypertension, SLE, IBD, Crohn’s disease, prior history of MHT, uterine fibroids, previous adnexal surgery, endometriosis, and CCI score. The subgroup analyses were performed based on age, and the survival analysis of thyroid cancer incidence between the two groups was performed. Additionally, we performed a sensitivity analysis focusing specifically on the widely used clinical measure CCI, which assesses various disorders and cancers 33 . Finally, despite existing studies, our study is the only one that comprehensively evaluated the association between hysterectomy and benign thyroid diseases, including hypothyroidism, hyperthyroidism, autoimmune thyroid disease, goiter, and non-toxic single thyroid nodule. Several limitations should be acknowledged. First, despite propensity score matching and adjustment for multiple covariates, residual confounding cannot be completely excluded inherent in observational database studies. Unmeasured or unknown factors (such as family history, dietary iodine intake, genetic factors, or radiation exposure) might not have been fully captured. By including uterine fibroids in propensity score matching, our matched cohort had a high fibroid prevalence (~ 74%), so study findings mainly reflect risks for women similar to hysterectomy candidates and may not fully apply to the general female population with lower fibroid rates. Adjusting for fibroids in regression instead could broaden the target group and change results due to model choices, but this needs future checks as current data limits prevent it. Second, our dataset did not include serum estrogen levels, and subgroup analysis by use of MHT was limited due to the small proportion of users and lack of detailed hormonal information. Data on hormonal status (premature ovarian insufficiency, polycystic ovarian syndrome and menstrual cycle irregularities), MHT usage, and indications for hysterectomy were limited or derived from claims data, which may be susceptible to misclassification. Third, increased healthcare utilization following hysterectomy may have introduced diagnostic bias: women undergoing hysterectomy may be more likely to receive frequent medical surveillance, especially in the first years after surgery. This could lead to higher detection rates of thyroid cancer or benign thyroid disease independent of true risk, potentially inflating the observed associations. Fourth, the study population consisted of South Korean women, a relatively homogeneous ethnic group with uniform access to national health screening and unique healthcare systems, as well as a high dietary iodine intake. Consequently, these factors may limit the generalizability of our findings to populations with different ethnic backgrounds, healthcare access, or varying levels of iodine intake. Finally, although thyroid outcomes were ascertained using diagnostic codes and linkage with health records, subclinical or unrecorded cases may have been missed. These findings suggest that women who undergo hysterectomy, even without concomitant oophorectomy, may be at increased risk for subsequent thyroid cancer and selected benign thyroid disorders. Clinicians should consider this potential association in counseling and long-term monitoring of hysterectomized women, particularly those with additional risk factors for thyroid disease. Further research is needed to elucidate the biological mechanisms underlying this relationship and to determine whether targeted surveillance or preventive strategies are warranted.

Hysterectomy is primarily performed for benign gynecological conditions such as fibroids and abnormal uterine bleeding, and is often accompanied by oophorectomy depending on underlying pathology and patient age 1 , 2 . This procedure induces changes in ovarian blood flow and hormonal milieu, with bilateral salpingo-oophorectomy resulting in abrupt estrogen deprivation 3 . There is growing interest in the interplay between female sex hormones and thyroid disease. Thyroid cancer incidence is 2–4 times higher in reproductive-age women than men, suggesting that sex-related hormonal factors may partly contribute to this disparity 4 , 5 . Similarly, benign thyroid nodules are more prevalent in women between puberty and menopause, and show slower growth after menopause 4 , 6 . The lifetime risk of developing thyroid nodules is strongly influenced by reproductive factors, such as age at menopause and duration of reproductive years 7 . However, age-related epidemiological trends alone cannot fully explain these observations, as many cancers, including colorectal cancer, also rise after age 40 irrespective of hormonal status 8 . Therefore, mechanistic evidence is critical to support the role of estrogens in thyroid disease pathogenesis. Recent experimental studies have demonstrated that both benign and malignant thyroid tissues express estrogen receptors, and that estrogen stimulation can promote thyrocyte proliferation and potentially tumorigenesis 9 – 11 . To date, epidemiological data on the association between hysterectomy and subsequent thyroid disease remain limited and sometimes conflicting 12 – 15 . In this study, we sought to evaluate the effect of hysterectomy, with and without adnexal surgery, on the risk of developing thyroid cancer and benign thyroid disorders in a large population-based cohort of South Korean women.

Below is the link to the electronic supplementary material. Supplementary Material 1 Supplementary Material 1 Supplementary Material 2 Supplementary Material 2 Supplementary Material 3 Supplementary Material 3