Abstract
Introduction
In recent years, dietary supplements have emerged as popular “natural” alternatives to conventional pharmacological treatments for various conditions, including endometriosis. The growing popularity of supplements for endometriosis-associated pain, fueled by an expanding and minimally regulated market, underscores the need for robust evidence of efficacy, as a prerequisite for any consideration on effectiveness. This meta-analysis synthesizes evidence from randomized, placebo-controlled trials (RCTs), the gold standard in evidence-based medicine, to assess the efficacy of dietary supplements in endometriosis-associated pain.
Methods
A systematic search of PubMed, Embase, Scopus, and the Cochrane Library was conducted up to November 5, 2024, in adherence to PRISMA 2020 guidelines. Two independent reviewers screened studies using PICOS criteria: reproductive-age women with endometriosis (Population), dietary supplements (Intervention), placebo (Comparator), and pain-related outcomes (Outcomes), assessed in placebo-controlled RCTs adhering to CONSORT standards (Study type). Three pain domains were evaluated: (i) symptom severity (visual analogue scale [VAS] for pelvic pain, dysmenorrhea, dyspareunia), (ii) pain catastrophizing, and (iii) quality of life, as measured by the Short Form-12 Health Survey (SF-12) and the Endometriosis Health Profile-30 (EHP-30). Risk of bias was assessed using the Cochrane RoB2 tool. Random-effects models were used to calculate pooled mean differences (MDs) and 95% confidence intervals (CIs). Statistical heterogeneity was assessed with the I2 statistic, and subgroup analyses explored clinically relevant confounders. Sensitivity analyses excluded studies with conflicts of interest or trustworthiness issues, as defined by the Obstetrics and Gynecology Editors’ Integrity Group (OGEIG). Publication bias was evaluated using Egger’s test, Begg’s test, and the trim-and-fill method. All analyses were conducted using STATA version 18.
Results
Nine RCTs (n = 545 subjects; 274 in the treatment group and 271 in the placebo group) were included. Only three met the “absolute” OGEIG trustworthiness criteria. No significant differences were observed between supplements and placebo for pelvic pain (pooled MD: −1.1; 95% CI: −3.0 to 0.8; I2 = 96.1%), dysmenorrhea (pooled MD: −2.0; 95% CI: −4.4 to 0.5; I2 = 93.8%), or dyspareunia (pooled MD: −2.0; 95% CI: −4.9 to 0.9; I2 = 96.5%). These findings remained consistent when the analysis was restricted to studies without conflicts of interest, those authored by researchers with no retractions, and those meeting OGEIG trustworthiness criteria. Subgroup analyses reduced heterogeneity and confirmed no significant benefits. Pain catastrophizing and quality-of-life measures showed little to no improvement.
Conclusion
While limited evidence precludes definitive conclusions about specific dietary supplements, available data suggest they lack efficacy for managing endometriosis-associated pain. Given the absence of demonstrated benefits, along with potential harms and costs, dietary supplements should not be recommended at this time for managing endometriosis-related pain.
Keywords: Dietary supplements, Endometriosis, Pelvic pain, Systematic review, Meta-analysis
Introduction
Endometriosis is a chronic, estrogen-dependent condition affecting approximately 190 million women worldwide [1]. Hormonal treatments that suppress ovulation and menstruation are universally recommended as first-line options in international guidelines [2, 3]. However, concerns about prolonged hormonal therapy, particularly in women over 35, have arisen due to potential long-term adverse effects, including a slight increase in breast cancer risk [4, 5], reported only by observational studies [6]. Amid growing concerns and increasing public discourse, many women are turning to alternative “natural” treatments, leading to a surge in the use of dietary supplements.
Globally, supplement use across all age groups has increased significantly over the past decade. Data from the Centers for Disease Control and Prevention (CDC) indicate that 57.6% of women aged 20 and older reported using dietary supplements in 2017–2018 [7]. Similarly, a 2022 survey conducted across 14 European countries revealed that nearly 90% of respondents reported using supplements, with women being the primary consumers [8].
While conventional pharmacological treatments are held to strict regulatory standards, rigorous clinical trials, and post-market surveillance, dietary supplements often enter the market with little to no evidence of clinical benefit [9]. This issue is particularly relevant for chronic conditions like endometriosis, which require long-term management. Robust evidence of efficacy from controlled settings is a critical prerequisite for evaluating their potential effectiveness in real-world practice and determining whether the use of these costly products is justified or should be discouraged.
This meta-analysis was conducted to provide a transparent and rigorous evaluation of the efficacy of dietary supplements in endometriosis-associated pain, aiming to guide healthcare providers in informed decision-making and empower women to make evidence-based health choices. Evidence was exclusively drawn from randomized controlled trials (RCTs), the gold standard for evidence-based medicine, and assessed using validated tools, including the recent Obstetrics and Gynecology Editors’ Integrity Group (OGEIG) instrument [10], to ensure methodological rigor and trustworthiness.
Methods
Study Design and Search Strategy
The study protocol was registered on PROSPERO (CRD42024607058). The findings were reported in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) 2020 guidelines [11]. Two independent reviewers (A.R., C.B.) conducted a comprehensive search of PubMed, Embase, Scopus, and Cochrane Library up to November 5, 2024. The search strategy utilized a combination of Medical Subject Headings (MeSH) terms and equivalent keywords (shown in online Appendix I; for all online suppl. material, see https://doi.org/10.1159/000545414).
Eligibility Criteria
Inclusion criteria were defined using the PICOS framework [12]: (1) Population: reproductive-age women with endometriosis diagnosed by surgery/histology or imaging (transvaginal ultrasound or magnetic resonance imaging); diagnoses based solely on medical records or self-reporting were excluded; (2) Intervention: dietary supplements assigned through randomization in a double- or triple-blinded design. As defined by the CDC, dietary supplements were considered as products taken orally (e.g., pills, capsules, tablets, or liquids) containing one or more dietary ingredients intended to supplement the diet [13]; (3) Comparison: placebo; (4) Outcomes: pain addressed via three pain domains: (i) pain symptom severity (pelvic pain, dysmenorrhea, dyspareunia) measured using the 100-mm visual analogue scale (VAS) or the 11-point numeric rating scale (NRS); (ii) pain catastrophizing assessed with the pain catastrophizing scale (PCS); (iii) health-related quality of life (HRQoL) measured using standardized tools, including the Short Form-12 Health Survey (SF-12) and the Endometriosis Health Profile-30 (EHP-30); (5) Study type: RCTs adhering to standards [14], published in peer-reviewed journals, available in English, approved by an institutional review board or ethics committee, and registered in publicly accessible databases.
Study Selection and Data Extraction
Records were screened independently by two reviewers (A.R., C.B.). Discrepancies were resolved by a third reviewer (N.S.). A full list of data abstracted is provided in online Appendix II.
Risk of Bias and Certainty of the Evidence Assessment
Risk of bias assessment was conducted independently by two reviewers (A.R., C.B.) using the revised Cochrane risk of bias tool for randomized trials (RoB2) [15]. Discrepancies were resolved by a third reviewer (N.S.). The certainty of the evidence was assessed using the Grading of Recommendations Assessment, Development and Evaluation (GRADE) approach [16].
Data Analysis
Means and standard deviations were obtained from individual studies, or calculated using appropriate formulas [17, 18]. Pooled mean differences (MDs) with 95% confidence intervals (CIs) were pooled using random-effects models following the DerSimonian-Laird method [19].
Statistical heterogeneity was assessed using I2 statistics, following Cochrane guidelines [20]: 0%–40% (not important), 30%–60% (moderate), 50%–90% (substantial), and 75%–100% (considerable).
On each outcome, leave-one-out meta-analyses were conducted to address the impact of each study on the estimate. For outcomes in which more studies were available, two additional sets of sensitivity analyses were conducted to restrict the evidence based on the trustworthiness of included RCTs: (i) excluding articles declaring conflict of interest with the pharmaceutical company producing the dietary supplement (checked on the manuscript and on the trial registry) or whose authors authored other papers on the topic retracted for plagiarism, ethical violations or because of concerns or issues about data or results (verified via retraction database [21]); (ii) excluding RCTs that did not met the trustworthiness criteria defined by Obstetrics and Gynecology Editors’ Integrity Group (OGEIG) [10].
Leave-one-out meta-analyses were performed to test result robustness. Sensitivity analyses were conducted to restrict the evidence based on methodological and ethical criteria, excluding: (i) Studies with conflicts of interest (COI) or authored by researchers associated with retracted publications on the topic due to plagiarism, ethical violations, or data integrity issues, verified through a retraction database [21]; (ii) RCTs not meeting trustworthiness criteria as defined by the OGEIG [10]. Subgroup analyses assessed clinically relevant confounders (≥2 RCTs per subgroup [22]), including treatment duration (≤60 vs. >60 days), supplement type, study location, baseline pain severity (VAS ≥3 vs. any pain), and concurrent hormonal or pain treatments. Univariate meta-regressions evaluated the relationship between supplementation duration (days) and pooled MDs.
Publication bias was assessed quantifying small-study effect with the linear regression-based method described by Egger et al. [23] and the adjusted rank correlation test proposed by Begg and Mazumdar [24]. Nonparametric trim-and-fill analysis of publication bias was implemented to provide a pooled estimate with 95% CIs based on the contribution of observed and imputed studies (imputing on the left). All analyses were performed using STATA version 18 (Stata Corp LLC, 2024, College Station, TX, USA).
Results
Studies Overview
Of 1,189 identified records, 679 were screened after duplicate removal. Two hundred and seventy-two were sought for retrieval, and 247 were assessed for eligibility. Nine double-blinded RCTs involving 545 women with endometriosis (274 in the treatment group vs. 271 in the placebo group), met eligibility criteria and were included [25–33]. The PRISMA 2020 flow diagram is shown in Figure 1. Study characteristics are summarized in Table 1. A summary of all meta-analysis results is provided in Table 2.
Fig. 1.
Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) 2020 flow diagram.
Table 1.
Overview of the randomized controlled trials (RCTs) included in the meta-analysis
| First author, year of publication | Study country | Sample size | Investigated treatment | Duration, weeks | Study population characteristics | Endometriosis assessment | Outcomes addresseda |
|---|---|---|---|---|---|---|---|
| a) treatment group | a) type of dietary supplement | a) total duration of the trial | a) pain severity at enrollment | a) diagnostic modality | |||
| b) placebo group | b) total dose | b) total duration of the treatment (dietary supplement vs. placebo) | b) concurrent HT during the study period | b) disease severity (r-AFS stage) or lesions localization | |||
| c) concurrent pain treatment during the study period | |||||||
| Almassinokiani et al. [25] (2016) | Iran | a) 19 | a) Vitamin D | a) 24 | a) VAS ≥3 | a) Laparoscopy | A1, A2 |
| b) 20 (analyzed 19) | b) 50,000 IU/week | b) 12 | b) No | b) Stage I, II, III, IV | |||
| c) NA | |||||||
| Amirsalari et al. [26] (2021) | Iran | a) 64 (analyzed 60) | a) Garlic | a), b) 12 | a) NA | a) Surgery | A1, A2, A3 |
| b) 67 (analyzed 60) | b) 400 mg/day (1,100 μg of allicin) | b) Yes | b) Stage I, II, III, IV | ||||
| c) Yes | |||||||
| Gudarzi et al. [27] (2024) | Iran | a) 34 (analyzed 34) | a) Curcumin | a), b) 8 | a) NA | a) Laparoscopy | A1, C2 |
| b) 34 (analyzed 34) | b) 1,000 mg/day | b) No | b) NA | ||||
| c) Yes | |||||||
| Khodaverdi et al. [28] (2019) | Iran | a) 19 (analyzed 16) | a) Lactobacillus acidophilus, Lactobacillus plantarum, Lactobacillus fermentum, Lactobacillus gasseri | a) 12 | a) VAS >4 | a) Laparoscopy | A1, A2, A3 |
| b) 18 (analyzed 16) | b) 109 colony/day | b) 8 | b) No | b) Stage III and IV | |||
| c) Yes | |||||||
| Mehdizadehkashi et al. [29] (2021) | Iran | a) 30 (analyzed 25) | a) Vitamin D | a), b) 12 | a) No | a) Surgery | A2, A3 |
| b) 30 (analyzed 25) | b) 50,000 IU/2 week | b) NA | b) NA | ||||
| c) NA | |||||||
| Mendes da Silva et al. [30] (2017) | Brazil | a) 22 (analyzed 22) | a) Resveratrol | a), b) 6 | a) NA | a) Laparoscopy | A1 |
| b) 22 (analyzed 22) | b) 40 mg/day | b) Yes | b) NA | ||||
| c) Yes | |||||||
| Mirzaei et al. [31] (2022) | Iran | a) 40 (analyzed 35) | a) Silymarin | a), b) 12 | a) Moderate to severe pain | a) 3D-TVUS | A1, C1 |
| b) 40 (analyzed 35) | b) 280 mg/day | b) Yes | b) all OMA | ||||
| c) NA | |||||||
| Nodler et al. [32] (2020) | USA | a) 27 (analyzed 24 in primary analysis; 22 and 23 in secondary analysis for quality of life and pain catastrophizing, respectively) | a) Vitamin D | a), b) 24 | a) VAS ≥3 | a) Surgery | B, C1 |
| b) 22 (analyzed 17 and 20 in primary analysis; 18 and 19 in secondary analysis for quality of life and pain catastrophizing, respectively) | b) 2,000 IU/day | b) Yes | b) Stage I, II, III, IV | ||||
| a) 20 (analyzed 19 in primary analysis; 16 and 17 in secondary analysis for quality of life and pain catastrophizing, respectively) | a) Fish oil | c) Yes | |||||
| b) 22 (analyzed 17 and 20 in primary analysis; 18 and 19 in secondary analysis for quality of life and pain catastrophizing, respectively) | b) 1,000 mg/day (720 ω-3 fatty acids, including 488 mg EPA and 178 mg DHA) | ||||||
| Söderman et al. [33] (2023) | Sweden | a) 20 (analyzed 20 in primary analysis; 17 and 18 in secondary analysis for dyspareunia and pain catastrophizing, respectively) | a) Melatonin | a), b) 8 | a) NRS >7 | a) Laparoscopy, TVUS, or MRI | A1, A2, B, C2 |
| b) 20 (analyzed 20 in primary analysis; 20 and 17 in secondary analysis for dyspareunia and pain catastrophizing, respectively) | b) 20 mg/day | b) Yes | b) NA | ||||
| c) Yes |
3D-TVUS, three-dimensional transvaginal ultrasonography; HT, hormonal treatment; MRI, magnetic resonance imaging; NA, not available; NRS, numerical rating scale; OMA, ovarian endometrioma; r-AFS, revised-American Fertility Society; TVUS, transvaginal ultrasound; VAS, visual analogue scale.
aOutcomes addressed categorized into three pain domains: A) pain symptom severity (VAS): A1 = pelvic Pain; A2 = dysmenorrhea; A3 = dyspareunia; B) pain catastrophizing: B = pain catastrophizing scale (PCS); C) health-related quality of life: C1 = Short Form Health Survey-12 (SF-12) (subscales: C1a = SF-12 – physical component summary [SF-12 – PCs]; C1b = SF-12 – mental component summary [SF-12 – MCs]); C2 = Endometriosis Health Profile-30 (EHP-30) (subscales: C2a = pain; C2b = control powerlessness; C2c = emotional and well-being; C2d = social support; C2e = self-image).
Table 2.
Summary of meta-analysis results
| Outcomes of the pain domains | Studies, n | Treatment vs. placebo, n | Effect estimate | Heterogeneity, I2 | |
|---|---|---|---|---|---|
| mean diff. | 95% CI | ||||
| A. Pain symptom severity | |||||
| A1. Pelvic pain | |||||
| Main analysis | 7 | 206 vs. 206 | −1.1 | −3.0 to 0.8 | 96.1% |
| Sensitivity analyses | |||||
| No COI/retractionsa | 6 | 190 vs. 190 | −1.2 | −3.3 to 0.9 | 96.7% |
| “Absolute” OGEIG trustworthiness criteriab | 3 | 102 vs. 102 | −2.0 | −5.4 to 1.4 | 96.9% |
| Subgroup analyses | |||||
| Study country | |||||
| Iran | 5 | 164 vs. 164 | −1.4 | −3.8 to 1.1 | 97.2% |
| Others | 2 | 42 vs. 42 | −0.5 | −1.4 to 0.4 | 0.0% |
| Pain severity at enrollment | |||||
| Any pain | 3 | 116 vs. 116 | −1.8 | −5.5 to 2.0 | 98.4% |
| At least moderate pain (VAS ≥3) | 4 | 90 vs. 90 | −0.6 | −1.5 to 0.2 | 50.3% |
| HT status | |||||
| No concurrent HT | 3 | 69 vs. 69 | 0.2 | −0.4 to 0.7 | 0.0% |
| Concurrent HT | 4 | 137 vs. 137 | −1.9 | −4.4 to 0.6 | 95.9% |
| Duration of the supplementation | |||||
| ≤60 days | 4 | 92 vs. 92 | −0.1 | −0.6 to 0.5 | 0.0% |
| >60 days | 3 | 114 vs. 114 | −2.2 | −5.4 to 1.0 | 97.3% |
| A2. Dysmenorrhea | |||||
| Main analysis | 4 | 120 vs. 120 | −2.0 | −4.4 to 0.5 | 93.8% |
| Sensitivity analyses | |||||
| No COI/retractionsa | 2 | 79 vs. 79 | −2.7 | −6.7 to 1.3 | 95.1% |
| “Absolute” OGEIG trustworthiness criteriab | 1 | NA | NA | NA | NA |
| Subgroup analyses | |||||
| Pain severity at enrollment | |||||
| Any pain | 2 | 85 vs. 85 | −2.8 | −6.6 to 1.0 | 96.5% |
| At least moderate pain (VAS ≥3) | 2 | 35 vs. 35 | −1.0 | −2.2 to 0.1 | 0.0% |
| HT status | |||||
| No concurrent HT | 2 | 35 vs. 35 | −1.0 | −2.2 to 0.1 | 0.0% |
| Concurrent HT | 1 | NA | NA | NA | NA |
| Duration of supplementation | |||||
| ≤60 days | 1 | NA | NA | NA | NA |
| >60 days | 3 | 104 vs. 104 | −2.1 | −5.2 to 0.9 | 95.3% |
| Type of supplement | |||||
| Vitamin D | 2 | 44 vs. 44 | −0.8 | −1.8 to 0.3 | 0.0% |
| A3. Dyspareunia | |||||
| Main analysis | 4 | 118 vs. 119 | −2.0 | −4.9 to 0.9 | 96.5% |
| Sensitivity analyses | |||||
| No COI/retractionsa/“absolute” OGEIG trustworthiness criteriab | 2 | 77 vs. 78 | −2.8 | −7.6 to 1.9 | 98.4% |
| Subgroup analyses | |||||
| Pain severity at enrollment | |||||
| Any pain | 2 | 85 vs. 85 | −3.4 | −7.2 to 0.4 | 96.1% |
| At least moderate pain (VAS ≥3) | 2 | 33 vs. 34 | −0.5 | −1.4 to 0.4 | 0.0% |
| Duration of supplementation | |||||
| ≤60 days | 2 | 33 vs. 34 | −0.5 | −1.4 to 0.4 | 0.0% |
| >60 days | 2 | 85 vs. 85 | −3.4 | −7.2 to 0.4 | 96.1% |
| B. Pain catastrophizing (by PCS) | |||||
| Main analysis | 3 (2)c | 61 vs. 56 | 0.3 | −2.0 to 2.6 | 17.3% |
| Sensitivity analyses | |||||
| No COI/retractionsa,d | 3 (2)c | 61 vs. 56 | 0.3 | −2.0 to 2.6 | 17.3% |
| “Absolute” OGEIG trustworthiness criteriab | 1 | NA | NA | NA | NA |
| C. Health-related quality of life | |||||
| C1. SF-12 | |||||
| C1a. SF-12 – Physical component summary (PCs) | |||||
| Main analysis | 3 (2)c | 78 vs. 69 | 1.2 | −6.1 to 8.6 | 88.3% |
| Sensitivity analyses | |||||
| No COI/retractionsa,d | 3 (2)c | 78 vs. 69 | 1.2 | −6.1 to 8.6 | 88.3% |
| “Absolute” OGEIG trustworthiness criteriab | 0 | NA | NA | NA | NA |
| C1b. SF-12 – Mental component summary (MCs) | |||||
| Main analysis | 3 (2)c | 78 vs. 69 | 3.6 | 0.8 to 6.3 | 0.0% |
| Sensitivity analyses | |||||
| No COI/retractionsa,d | 3 (2)c | 78 vs. 69 | 3.6 | 0.8 to 6.3 | 0.0% |
| “Absolute” OGEIG trustworthiness criteriab | 0 | NA | NA | NA | NA |
| C2. EHP-30 | |||||
| C2a. Pain | |||||
| Main analysis | 2 | 52 vs. 50 | 6.7 | 0.0 to 13.4 | 0.0% |
| Sensitivity analyses | |||||
| No COI/retractionsa,d | 2 | 52 vs. 50 | 6.7 | 0.0 to 13.4 | 0.0% |
| “Absolute” OGEIG trustworthiness criteriab | 1 | NA | NA | NA | NA |
| C2b. Control and powerlessness | |||||
| Main analysis | 2 | 52 vs. 50 | 3.8 | −3.9 to 11.5 | 0.0% |
| Sensitivity analyses | |||||
| No COI/retractionsa,d | 2 | 52 vs. 50 | 3.8 | −3.9 to 11.5 | 0.0% |
| “Absolute” OGEIG trustworthiness criteriab | 1 | NA | NA | NA | NA |
| C2c. Emotional well-being | |||||
| Main analysis | 2 | 52 vs. 50 | 0.3 | −6.9 to 7.6 | 0.0% |
| Sensitivity analyses | |||||
| No COI/retractionsa,d | 2 | 52 vs. 50 | 0.3 | −6.9 to 7.6 | 0.0% |
| “Absolute” OGEIG trustworthiness criteriab | 1 | NA | NA | NA | NA |
| C2d. Social support | |||||
| Main analysis | 2 | 52 vs. 50 | 0.7 | −7.4 to 8.8 | 0.0% |
| Sensitivity analyses | |||||
| No COI/retractionsa,d | 2 | 52 vs. 50 | 0.7 | −7.4 to 8.8 | 0.0% |
| “Absolute” OGEIG trustworthiness criteriab | 1 | NA | NA | NA | NA |
| C2e. Self-image | |||||
| Main analysis | 2 | 52 vs. 50 | 1.9 | −5.9 to 9.6 | 0.0% |
| Sensitivity analyses | |||||
| No COI/retractionsa,d | 2 | 52 vs. 50 | 1.9 | −5.9 to 9.6 | 0.0% |
| “Absolute” OGEIG trustworthiness criteriab | 1 | NA | NA | NA | NA |
95% CI, 95% confidence interval; COI, conflicts of interest; EHP-30, Endometriosis Health Profile-30; HT, hormonal treatment; mean diff., mean difference; MCs, mental component summary; NA, not available; OGEIG, Obstetrics and Gynecology Editors’ Integrity Group; PCS, pain catastrophizing scale; PCs, physical component summary; SF-12, Short Form-12 Health Survey; VAS, visual analogue scale.
aCOI with the pharmaceutical company producing the dietary supplement checked on the manuscript and on the trial registries; retractions for plagiarism, ethical violations or because of concerns or issues about data or results checked on retraction database [21]. Full assessment reported in Supplement Table 1.
b“Absolute” OGEIG trustworthiness criteria according OGEIG, 2024 [10]. Full assessment reported in online supplementary Table 2 and 3.
cTwo of the three available comparisons between supplements and placebo came from the same multi-arm parallel RCT.
dNone of the authors of included studies reported any COI or authored retracted works on the topic.
Risk of Bias and Trustworthiness Assessment
Risk of bias (RoB2) [15] is detailed in online supplementary Figure 1. Of the nine included RCTs, one [28] was funded by the supplement manufacturer, and another [29] was authored by an author with 14 retracted manuscripts on the topic due to ethical violations or data concerns (shown in online suppl. Table 1).
Most of the included studies had significant methodological concerns, as outlined in online supplementary Table 2. Specifically, one study was registered during the recruitment phase [25], and two were registered retrospectively [28, 29]. Only three RCTs [26, 30, 33] met the “Absolute” trustworthiness criteria defined by OGEIG [10] (see online suppl. Table 3), with one of them [26] showing baseline differences between groups, raising questions about the validity of the randomization process.
Pain Symptom Severity
Pelvic Pain
The pooled mean difference (MD) in VAS for pelvic pain showed no significant difference between treatment and placebo (−1.1; 95% CI: −3.0 to 0.8; 7 studies; n = 412), with high heterogeneity (I2 = 96.1%), and no publication bias (Egger’s p = 0.31, z = 1.0; Begg’s p = 0.76, z = −0.6; shown in Fig. 2). The certainty of evidence was graded as moderate according to GRADE [16] (shown in online suppl. Table 4). Excluding Amirsalari et al. [26] reduced heterogeneity (I2 = 50.0%) and resulted in a pooled MD of −0.4 (95% CI: −1.1 to 0.2) (shown in online suppl. Fig. 2). Sensitivity analyses confirmed no significant benefit, irrespective of COI, or the application of OGEIG trustworthiness criteria [10] (shown in online suppl. Fig. 3).
Fig. 2.
Pooled mean difference in pelvic pain severity by visual analogue scale (VAS) in treatment versus placebo. 95% CI, 95% confidence interval; mean diff., mean difference; SD, standard deviation.
Subgroup analyses showed no treatment effect versus placebo, with no heterogeneity (I2 = 0.0%) when considering studies conducted outside Iran, those without concurrent HT, or supplementation duration <60 days (shown in online suppl. Fig. 4). Random-effects meta-regression confirmed no significant impact of treatment duration on pooled MD (coefficient −0.1; 95% CI: −0.1 to 0.0; p = 0.15).
Dysmenorrhea
The pooled MD in VAS for dysmenorrhea showed no significant difference between treatment and placebo (−2.0; 95% CI: −4.4 to 0.5; 4 studies; n = 240), with high heterogeneity (I2 = 93.8%) (shown in Fig. 3). Evidence of publication bias was detected (Egger’s p < 0.001, z = 3.5; Begg’s p = 0.73, z = 0.3), and the trim-and-fill method adjusted for one study, yielding a pooled MD of −2.4 (95% CI: −4.4 to −0.4) (shown in online suppl. Fig. 5). The certainty of evidence was graded as moderate-to-low according to GRADE [16] (shown in online suppl. Table 4). Excluding Amirsalari et al. [26] in leave-one-out analysis removed heterogeneity (I2 = 0.0%) and revealed a significant but modest difference in treatment versus placebo (−0.9; 95% CI: −1.8 to −0.1; p = 0.03; shown in online suppl. Fig. 6). Sensitivity analyses confirmed no significant benefit, regardless of conflicts of COI or authors’ retractions on the topic (shown in online suppl. Fig. 7).
Fig. 3.
Pooled mean difference in dysmenorrhea severity by visual analogue scale (VAS) in treatment versus placebo. 95% CI, 95% confidence interval; mean diff., mean difference; SD, standard deviation.
Subgroup analyses showed no treatment effect versus placebo, with no heterogeneity (I2 = 0.0%) in studies without concurrent HT or enrolling only women with VAS ≥3 (shown in online suppl. Fig. 8). Two studies [25, 29] using vitamin D as supplementation provided consistent results (shown in online suppl. Fig. 8). No difference was observed in subgroup analysis by supplementation duration, and random-effects meta-regression confirmed no significant impact of treatment duration on pooled MD (coefficient −0.1; 95% CI: −0.2 to 0.1; p = 0.62).
Dyspareunia
The pooled MD in VAS for dyspareunia showed no significant difference between treatment and placebo (−2.0; 95% CI: −4.9 to 0.9; 4 studies; n = 237), with high heterogeneity (I2 = 96.5%) and no publication bias (Egger’s p = 0.17, z = 1.4; Begg’s p = 0.73, z = 0.3; shown in online suppl. Fig. 9). The certainty of evidence was graded as moderate-to-low according to GRADE [16] (shown in online suppl. Table 4). Excluding Amirsalari et al. [26] yielded a more precise estimate, with a modest reduction in pooled MD (−0.8; 95% CI: −1.5 to 0.0; p = 0.05) and no heterogeneity (I2 = 0.0%) (shown in online suppl. Fig. 10). Sensitivity analyses based on OGEIG trustworthiness criteria [10] also confirmed no significant benefit (shown in online suppl. Fig. 11).
Subgroup analyses showed no treatment effect versus placebo, with no heterogeneity (I2 = 0.0%) in studies enrolling women with VAS ≥3 (shown in online suppl. Fig. 12). No difference was found in subgroup analysis by supplementation duration, and random-effects meta-regression confirmed no significant impact of treatment duration on pooled MD (coefficient −0.1; 95% CI: −0.2 to 0.0; p = 0.62).
Pain Catastrophizing
The pooled MD in pain catastrophizing scale (PCS) showed no significant difference between treatment and placebo (0.3; 95% CI: −4.9 to 0.9; 2 studies, 3 comparisons; n = 117), with low heterogeneity (I2 = 17.3%) and no publication bias (Egger’s p = 0.96; z = −0.1; Begg’s p = 1.00; z = −1.0; shown in online suppl. Fig. 13). None of the included RCTs reported any COI, and no authors had retracted works on the topic. Only one study [33] met OGEIG trustworthiness criteria [10].
Health-Related Quality of Life
Short Form-12 Health Survey
The pooled MDs in SF-12 – physical component summary (SF-12 – PCs) and SF-12 – mental component summary (SF-12 – MCs) in treatment vs. placebo (2 studies [32, 33], 3 comparisons; n = 147) were 1.2 (95% CI: −6.1 to 8.6; I2 = 88.3%) and 3.6 (95% CI: 0.8 to 6.3; I2 = 0.0%), respectively. No publication bias was detected (SF-12 – PCs: Egger’s p = 0.39, z = 0.9; Begg’s p = 1.00, z = 0.0; SF-12 – MCs: Egger’s p = 0.27, z = −1.1; Begg’s p = 0.29; z = −2.1; shown in online suppl. Fig. 14). None of the included RCTs reported COI or were authored by researchers with retractions on the topic. None met OGEIG trustworthiness criteria [10].
Endometriosis Health Profile-30
The pooled MDs in EHP domains at study end (2 studies [28, 34]; n = 102) were not significant in any of the 5 domains (pain: pooled MD 6.7, 95% CI: 0.0 to 13.4; control and powerlessness: pooled MD 3.8, 95% CI: −3.9 to 11.5; emotional well-being: pooled MD: 0.3, 95% CI: −6.9 to 7.6; social support: pooled MD: 0.7, 95% CI: −7.4 to 8.8; self-image: pooled MD: 1.9, 95% CI: −5.6 to 9.6), with no heterogeneity (I2 = 0.0% in all pooled estimates) and nonsignificant publication bias (pain: Egger’s p = 0.73, z = −0.4; control and powerlessness: Egger’s p = 0.64, z = −0.5; emotional and well-being: Egger’s p = 0.66, z = −0.4; social support: Egger’s p = 0.75, z = −0.3; self-image: Egger’s p = 0.97, z = 0.0; shown in online suppl. Fig. 15). None of the included RCTs reported COI or were authored by researchers with retractions on the topic, and only one [33] met OGEIG trustworthiness criteria [10].
Discussion
Main Findings
This meta-analysis reveals that most RCTs investigating dietary supplements for managing endometriosis-associated pain have significant methodological issues and cannot be considered trustworthy. Based on the available evidence, this analysis found no significant efficacy of dietary supplements over placebo across various pain-related domains, including symptom severity, pain catastrophizing, and HRQoL.
For pain symptom severity, no differences were observed in VAS scores for pelvic pain or dyspareunia, with no evidence of publication bias despite high heterogeneity. Similarly, no overall benefit was observed for dysmenorrhea. However, high heterogeneity and publication bias prevented ruling out some, inconsistent efficacy in sensitivity analyses. Importantly, in studies excluding concurrent HT – considered the most appropriate setting for evaluating pain during menstruation – consistent findings of no benefit were observed. The lack of efficacy across all pain symptoms remained consistent when analyses were restricted to studies without COI or authors’ previous retractions on the topic and to RCTs meeting OGEIG trustworthiness criteria [10]. Subgroup analyses further confirmed no treatment effects for pain symptom severity, with low or no heterogeneity in studies conducted outside Iran, those enrolling women with moderate or greater pain at baseline, and those excluding concurrent HT. Additionally, the duration of supplementation did not influence outcomes, whether assessed as a dichotomous variable (≤60 vs. >60 days) or as a continuous measure (in days).
Regarding HRQoL, no differences were observed in the SF-12 – PCs, while a minimal benefit was detected in the SF-12 – PCS, despite the limited number of studies. Disease-specific quality-of-life measures assessed using the EHP-30 demonstrated no significant differences across any of the five evaluated domains. Specific evidence for individual supplements was insufficient for pooling, apart from vitamin D for dysmenorrhea, which also showed no efficacy compared to placebo.
Interpretation
The dietary supplements market is experiencing significant growth worldwide. In Europe, it was valued at USD 14.95 billion in 2019 and is projected to reach USD 33.80 billion by 2027, with a compound annual growth rate of 9.3% from 2020 to 2027 [34]. Similarly, in the USA, the market was valued at USD 53.58 billion in 2023 and is expected to grow at a compound annual growth rate of 5.7% between 2024 and 2030 [35].
In a rapidly expanding global market, healthcare providers and patients are faced with an overwhelming array of dietary supplement options. However, in line with the principles of evidence-based medicine [36], recommending any compound, including dietary supplements, for effectiveness in real-world settings requires robust evidence of efficacy from reliable, evidence-based sources. As an example, robust data from large-scale RCTs demonstrating the unequivocal and significant benefits of daily folic acid in preventing neural tube defects in offspring have established it as a worldwide cornerstone of preconception care [37, 38].
Regarding endometriosis, some authors have proposed the use of nutritional supplements, either alone or in combination with other interventions, as a potential approach for managing endometriosis-related symptoms, based on the anti-inflammatory and antioxidant properties of various molecules [39, 40]. However, the findings of this meta-analysis, which includes only evidence-based data from RCTs, highlight the lack of robust evidence supporting the use of dietary supplements for managing endometriosis-associated pain or improving related outcomes. This is consistent with general evidence discouraging the routine use of dietary supplementation in chronic diseases due to insufficient benefits and a lack of demonstrated impact on prevention or treatment [41]. In fact, in patients with chronic diseases undergoing long-term pharmacological therapy, the use of dietary supplements should be carefully evaluated, considering potential interactions, the risk of excessive intake, and additional costs, particularly in the absence of robust clinical evidence supporting their efficacy [42].
A critical finding from our meta-analysis is that only three RCTs [26, 30, 33] met the rigorous methodological and ethical standards outlined by the OGEIG trustworthiness criteria [10]. This underscores a broader issue in the field of dietary supplements: these products often enter the market with minimal regulatory oversight, unlike traditional pharmacological treatments. Moreover, the lack of harmonization in dietary supplement regulations highlights the absence of unified global standards, leaving regulatory obligations for both authorities and manufacturers inadequately defined, particularly regarding product efficacy and safety [43].
In the USA, the Food and Drug Administration (FDA) requires manufacturers to evaluate the identity, purity, strength, and composition of their products. However, these regulations do not mandate pre-market evidence of safety, and manufacturers are not required to submit such data before marketing their products. This reliance on post-market monitoring and retrospective labeling actions has repeatedly proven inadequate, as highlighted by numerous reports exposing systemic flaws [44]. The regulatory gap increases the risk of contamination or mislabeling and undermines the overall integrity of marketed dietary supplements. Contamination of dietary supplements is a particularly alarming issue, with studies revealing the presence of undeclared pharmaceutical agents, heavy metals, and microbial contaminants in products marketed as “safe” and “natural.” For instance, a systematic review found that up to 15% of supplements marketed for athletes contained undeclared anabolic steroids or stimulants [45]. Similarly, microbial contamination has been detected in herbal supplements, including harmful bacteria and molds [46]. Heavy metal contamination is another widespread concern, with high levels of lead, cadmium, and mercury identified in numerous dietary supplements, posing significant health risks to consumers [47].
Furthermore, the lack of pre-marketing standardized dosing guidelines for dietary supplements exposes consumers to significant risks associated with excessive consumption [48]. Nutrients play specific biochemical and physiological roles in the body, with systemic homeostasis maintained through selective mechanisms. Overconsumption of nutrients, such as supplementation without need, can disrupt this delicate balance and lead to harm. To mitigate risks associated with excessive supplement consumption, the European Food Safety Authority (EFSA) is currently working to develop guidance on micronutrient bioavailability and tolerable upper intake levels [49, 50].
According to a recent large population-based study, the most common reasons for supplement use are to “improve” (45%) or “maintain” (33%) health [51]. However, data from the National Health and Nutrition Examination Survey (NHANES) indicate that in the pre-pandemic era, approximately 85% of individuals who took dietary supplements for perceived immune benefits did so without a healthcare professional’s recommendation, and 60% relied solely on label claims regarding immune benefits [52]. Therefore, misinformation remains, to date, a primary driver for patients choosing “alternative” treatments as adjuncts or substitutes for evidence-based care [53]. Conversely, to steer consumer behavior toward demonstrable health benefits and prioritize value-based healthcare, patient information must be firmly grounded in evidence-based medicine, ensuring optimized quality of care and efficient resource allocation.
Strengths and Limitations
The primary strength of this study is that it provides the first comprehensive analysis of the effect of dietary supplements on multiple pain domains in endometriosis, offering a comprehensive evaluation using the most widely applied patient-reported outcomes in clinical practice. The consistency of effects across multiple pain domains, with most pooled estimates nearing the no-effect line, reinforces the strength of our findings. Another key strength is the exclusion of all non-randomized studies, as relying on such studies fails to adequately control for potential biases and introduces additional uncertainty regarding therapeutic effects, particularly for newer potential treatments like supplements [54]. Notably, the analysis of trustworthiness is a key strength of this study, as it highlights that most of the available evidence is of poor quality, reinforcing the conclusion that there is insufficient evidence to support the efficacy of these interventions.
The main limitation of this meta-analysis is the limited number of available RCTs, which prevented the calculation of effect estimates specific to individual dietary supplements. However, primary studies generally showed consistent effect estimates, with one exception [26]. To account for variability, we employed a pooling method [20] designed to address both within-study sampling error and between-study heterogeneity, ensuring a more realistic overall effect size when studies are not entirely homogeneous. Another limitation is that, although planned a priori, subgroup analyses based on optimized concurrent pain treatments could not be conducted, as all studies included in the analysis implemented optimized pain management during the study period. As a result, in the absence of mediation analysis, the observed effects of dietary supplements in the included RCTs are likely partially influenced by concurrent pain and hormonal treatments.
Conclusions
The use of dietary supplements within evidence-based medicine remains challenging due to insufficient robust evidence and the lack of standardized guidelines. While supplements may help meet nutrient needs, they also pose risks of excessive intake. Our findings do not support their use for managing endometriosis-associated pain, showing no efficacy in reducing pain severity, pain catastrophizing, or improving quality of life. Unless specific nutritional deficiencies are identified, supplements should not be recommended due to their cost, lack of efficacy, and potential harms.
For chronic conditions like endometriosis, supporting self-management is key to alleviating pain and fatigue while addressing broader impacts on quality of life [55]. Effective self-management relies on shared decision-making between patients and healthcare providers [56], backed by evidence-based, accessible information, including insights into “natural” products. Improved communication and evidence-based discussions are essential to guide patients toward safe, effective treatments and to ensure they understand the potential harms and lack of proven benefits of dietary supplements as alternatives to hormonal therapies.
Statement of Ethics
A Statement of Ethics is not applicable because this study is based exclusively on published literature.
Conflict of Interest Statement
P.Ve. is a member of the Editorial Board of Human Reproduction Open and the Journal of Obstetrics and Gynaecology Canada and the International Editorial Board of Acta Obstetricia et Gynecologica Scandinavica; has received royalties from Wolters Kluwer for chapters on endometriosis management in the clinical decision support resource UpToDate; and maintains both a public and private gynecologic practice. P.Vi. is the Co-Editor-in-Chief of the Journal of Endometriosis and Uterine Disorders. E.S. is Editor-in-Chief of Human Reproduction Open; discloses payments from Ferring for research grants; and receives honoraria from Institut Biochimique Société Anonyme (IBSA) and Gedeon-Richter for lectures, and maintains both a public and private gynecologic practice. No other disclosures were reported.
Funding Sources
This study was partially supported by the Italian Ministry of Health – Current Research IRCCS.
Author Contributions
N.S.: conceptualization, methodology, software, validation, formal analysis, investigation, resources, data curation, writing – original draft preparation, visualization, and project administration. A.R.: validation, data curation, visualization, and resources. C.B.: validation, data curation, visualization, and resources. E.S. and P.Vi.: validation and writing – review and editing. P.Ve.: validation, resources, writing – review and editing, and supervision.
Funding Statement
This study was partially supported by the Italian Ministry of Health – Current Research IRCCS.
Data Availability Statement
All data generated or analyzed during this study are included in this article and its supplementary material files. Further inquiries can be directed to the corresponding author.
Supplementary Material.
Supplementary Material.
Supplementary Material.
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Associated Data
This section collects any data citations, data availability statements, or supplementary materials included in this article.
Supplementary Materials
Data Availability Statement
All data generated or analyzed during this study are included in this article and its supplementary material files. Further inquiries can be directed to the corresponding author.
