The quality of evidence was very low, and triangulation was of limited utility because alternative methods were used infrequently and often not robustly applied.Prenatal nutrition is associated with offspring autism spectrum disorder (herein referred to as autism), yet, it remains unknown if the association is causal. Triangulation may improve causal inference by integrating the results of conventional multivariate regression with several alternative approaches that have unrelated sources of bias. We systematically reviewed the literature on the relationship between prenatal multivitamin supplements and offspring autism, and evidence for the causal approaches applied. Six databases were searched up to 8 June 2020, by which time we had screened 1309 titles/abstracts, and retained 12 articles. Quality assessment was guided using Newcastle–Ottawa in individual studies, and the Grading of Recommendations Assessment, Development and Evaluation (GRADE) for the body of evidence. The effect estimates from multivariate regression were meta-analysed in a random effects model and causal approaches were narratively synthesised. The meta-analysis of prenatal multivitamin supplements involved 904,947 children (8159 cases), and in the overall analysis showed no robust association with offspring autism; however, a reduced risk was observed in the subgroup of high-quality observational studies , I Autism spectrum disorder (hereafter “autism”) is a neurodevelopmental condition characterised by early-onset impairment in social communication and restricted and repetitive behaviour . PrenataPrevious reviews and meta-analysis reported a reduced risk of offspring autism associated with prenatal folic acid (FA) or multivitamin supplements ,5, but aAetiological triangulation recognises that all approaches have bias and then exploits unrelated sources of bias to “test” the consistency of results . Causal Evidence from alternative causal approaches is unsystematically synthesised in previous reviews, if at all, which limits transparency. As alternative causal approaches are increasingly applied within studies and may significantly alter causal reasoning, this evidence should be integrated in an explicit and scientifically rigorous process. This aligns with guidance from Cochrane  and a rehttps://www.crd.york.ac.uk/prospero (accessed on 23 July 2021). The Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) were followed  Supplem. The revThe study designs were trials, cohort, case-controls and cross-sectional studies of any duration. We focused on the use of folic acid and multivitamin supplements (hereafter simply “multivitamin supplement”) in women during preconception and prenatal periods because folic acid is generally sourced from a multivitamin supplement . ComparaThe search strategy and selection of databases were guided by an information scientist with expertise in systematic reviews. Search strategies were adapted to each database. The following databases were searched from the earliest date to 8 June 2020; MEDLINE (OVID), EMBASE (OVID), PsycINFO (EBSCO), Web of Science core collection, Open Grey and BioRix. See Titles and abstracts were screened, full articles reviewed, and quality assessment was completed twice independently for each study, by C.F., T.B., and M.S. Disagreements were resolved through discussion and adjudicated, by J.A. Data were extracted by C.F. using a standardised form comprised of, author, year of publication, country and cohort, study design, sample size, age of participants, nutritional supplement, measure of autism, covariates, results, and causal approach.The Newcastle–Ottawa Scale guided the quality assessment of each observational study. Scores range from 0 to 9, where a score of 7–9 was considered high quality in the subgroup analysis and consistent with similar previous reviews ,12. The We narratively synthesised studies that were inappropriate to meta-analysis. Additionally, we summarised the alternative approaches applied in table format.2 statistics. The interpretation of heterogeneity (I2) was guided by Cochrane’s reference ranges [We conducted a meta-analysis of the fully adjusted effect estimates in a random effects model using the Hartung–Knapp–Sidik–Jonkman estimator . Analysee ranges . Predicte ranges ,14,15. Te ranges . Statistp < 0.1 (2-sided). Small-study publication bias was assessed through the inspection of the funnel plot and Egger’s test [Sources of heterogeneity were explored through the identification of outliers, leave-one-out analysis and subgroup analysis if there were ≥10 studies. A random effects model was used to estimate between and within subgroup effects. The pre-defined subgroups were study quality, study design (prospective/retrospective), region, mandatory fortification (yes/no), and stage of pregnancy, which was defined as the first trimester, compared against any point in pregnancy, if the nutritional supplement exposure period was undefined. Between-subgroup differences were measured with the q statistic and significance was indicated by r’s test . The widr’s test , but to A total of 1309 titles were identified with 342 duplicates, leaving 967 titles and abstracts to be screened . Of thesBased on the Newcastle–Ottawa Scale, five ,23,24,25A total of 904,947 children including 8159 cases from six countries were included in the meta-analysis. All studies measured nutritional supplements and one included fortified food. The multivitamin dose was seldom reported.2 = 94.3%, p < 0.001). Egger’s test (p = 0.44) and inspection of the funnel plot suggests no evidence of asymmetry (2 = 94.3%). Upon removal of the outlier, DeSoto and Hitlan, there was a 32% reduced risk of autism ; although heterogeneity remained considerable. There were no other influential studies  . The consymmetry . The pre studies . The 95%2 = 62.4%). No association was observed in low-quality studies , and there were no between-subgroup differences detected Q = 0.00, p = 0.98). Some evidence of association was observed in prospective studies , whilst no association was evident in the retrospective studies . There were no differences between subgroups for study design . Subgroup analysis of regions reduced heterogeneity from “considerable” in Nordic  and American  studies. Asian countries had the largest effect estimate, but there was considerable heterogeneity . No differences between subgroups were detected . Subgroup analysis of the stage of pregnancy showed similar effect estimates in each group . There were no between-subgroup differences . Regions without mandatory fortification produced a stronger reduction in the risk of autism , compared to regions with mandatory fortification . There were no between-subgroup differences detected .See All studies ,28,29,30Prenatal multivitamin supplements were not robustly associated with autism in the overall meta-analysis. However, a reduced risk of autism was observed in high-quality studies, prospective studies, early pregnancy and following the removal of an outlier. In contrast with previous meta-analyses, we did not observe any strong evidence of association in our main results  due to the selection of the Hartung–Knapp–Sidik–Jonkman estimator, rather than DerSimonian–Laird which underestimates uncertainty and was applied in previous meta-analysis ,5,31. WeRegional variation in baseline nutritional status and genotype ,33 may bFurthermore, from the reviewed studies, only the U.S. implements mandatory fortification of diet with folic acid and so has a high baseline folic acid intake. Less than 1% of its population has had deficient plasma folate levels since the introduction of mandatory fortification in 1998 , but thiThe two studies in this review that observed a reduced risk of autism had folic acid intakes well in excess of the recommended 400 ug/day ,28, yet An additional source of heterogeneity is variation in the stage of pregnancy when the multivitamin supplement was taken. In early pregnancy it was associated with a lower risk of autism and lower heterogeneity. Periconception is an established critical period in which FA can reduce the incidence of neural tube defects by up to 70% . FurtherLastly, autism is a heterogeneous condition and specific features might have differential causal pathways , which mAlternative causal approaches were infrequently applied, and the application was often not robust; therefore, our ability to triangulate the findings was limited. All studies used multivariate regression, a key assumption of which is no residual confounding . HoweverIn Schmidt et al., an association between folate use and autism risk was only observed if the mother/child had the methylenetetrahydrofolate reductase (MTHFR) 677 C > T genotype, which is unlikely owing to confounding by socioeconomic and lifestyle characteristics . The MTHDiscordant sibling analyses may overcome shared unmeasured confounding despite several key methodological considerations. In DeVilbiss et al., the reduced risk of autism observed in their main analysis was attenuated by the discordant sibling analysis . SiblingIn two studies, negative controls were applied and acted as mock exposures that indicate the presence of bias without relying on the assumption of no unmeasured confounding . InsteadSuren et al. measured an association with multivitamin supplements, but not fish oils  even thoCollectively, triangulation as a strategy to further causal interpretation was limited, many due to infrequent use and limitations in applying alternative approaches. The multivariate regression and gene-nutrient interaction  findingsThis review has several strengths. Numerous steps to reduce bias were taken, such as searching grey literature sources, applying GRADE guidelines, calculating prediction intervals and using the Hartung–Knapp–Sidik–Jonkman estimator. However, the strongest advance was a formal narrative synthesis of the range of causal approaches. As an explicit approach, it provided transparent evidence of the approaches applied, their findings and outlined areas for future studies. Nonetheless, there were weaknesses, mainly the heterogeneity and inconsistency observed across studies and the low study numbers. The prediction intervals indicated if there had been an effect, it could have ranged from beneficial to null or even harmful in individual settings.At present, the evidence is inconclusive, and we are unable to confirm a causal association between prenatal multivitamin supplement use and autism in offspring. Future studies should improve the study design and data analyses through adequately powered prospective birth cohorts. The measurement of nutritional supplements could be improved through reporting their nutrient composition, dose, compliance and duration and timing of use, which are all known to affect biological responses but were rarely considered in the studies we reviewed. Furthermore, there should be greater consideration of the complexity of nutrition by modelling U-shaped relationships and considering how the response to nutrients is altered by variations in baseline requirements and whether it is affected by recent nutrient intake, changes to physiological demand in early pregnancy, or genetic variation. Lastly, we recommend the application of alternative approaches within a triangulation framework to gauge causality better.


The association between adiposity and papillary thyroid carcinoma (PTC) has been reported in several studies, but its association with aggressive clinicopathologic features is not well-recognized. Our aim is to systematically review the literature to identify whether adiposity, expressed through Body Mass Index (BMI), is related to aggressive clinicopathologic features such as tumor-node-metastasis (TNM) stage, extrathyroidal extension (ETE), lymph node (LN) metastasis and multifocality in patients with PTC.A systematic search for articles was performed using the PubMed, EBSCO, and Cochrane Library for all articles published in English until December 2020. Specific keywords such as “papillary thyroid carcinoma”, “Body Mass Index”, “clinicopathologic features” were used in the search strategy. Two independent reviewers screened all retrieved articles based on predefined inclusion and exclusion criteria. Meta-analysis was performed in the studies that reported crude and adjusted odds ratios (OR). The methodological quality was assessed using the Newcastle-Ottawa Scale.A total of 11 retrospective cohort studies involving 26,196 participants included. Our findings showed that elevated BMI was significantly associated with ETE in both overweight  and obesity group . Elevated BMI was also significantly associated with multifocality in overweight patients  and obese patients . Also, obesity was significantly associated with increased tumor size  and with LN metastasis , whereas being overweight was significantly associated with advanced TNM stage  Our results provide strong evidence for the association between higher BMI and ETE, multifocality, and tumor size. Further studies with a larger number of participants are required to elucidate further the association of increased BMI with advanced TNM stage and LN metastasis. The prevalence of thyroid carcinoma has significantly increased worldwide over the past decades, with the highest percentage of increase attributed to a rise in the incidence of papillary thyroid carcinoma (PTC) and predominantly small PTC –3. ThyroSeveral environmental factors and eating habits, such as iodine intake, cruciferous vegetable consumption, fish consumption, obesity, radiation exposure, imaging with iodine-containing media, and endocrine disruptors, have been implicated in the pathogenesis of thyroid cancer increase , 10. ObeA definitive association of adiposity and the development of thyroid cancer has not been well established yet. Still, epidemiological data are pointing to an independent association between obesity and increased incidence of thyroid neoplasia. A meta-analysis including 12,199 thyroid cases reported a 25% greater risk of thyroid cancer in overweight patients and a 55% increase in thyroid cancer risk among obese patients when compared with normal weighted individuals . VariousThis systematic review and meta-analysis aimed to examine the association between BMI and aggressive clinicopathologic features such as tumour-node-metastasis (TNM) stage, extrathyroidal extension (ETE), lymph node (LN) metastasis, and multifocality in patients with PTC. To our knowledge, there are no previous systematic reviews or meta-analyses that evaluated the association of BMI with aggressive clinicopathologic features in patients with PTC. Evidence of such associations would be extremely important in daily clinical practice, as weight loss programs could become part of PTC patient management.We followed the Preferred Reporting Items for Systematic Reviews and Meta-Analyses guidelines (PRISMA) . The PRIWe used the Participants, Interventions, Comparisons, Outcomes, and Study design (PICOS) approach  to identTwo authors (AE & KG) extracted the data independently, and any discrepancy was resolved after consultation with a third author (PP). The extracted data collected included: first author, year of publication, country of origin, design of the study, sample size, characteristics of the study population  and reported odds ratios (OR) (both adjusted and unadjusted) or rates of tumours size, ETE, multifocality, LN metastasis, and TNM stage. The methodological quality was performed using the Newcastle- Ottawa Scale (NOS) . The NOS2-based Cochran Q test. The variation in estimates attributable to heterogeneity was quantified by the measure I2 metric for inconsistency (2 index (>50% indicating significant heterogeneity). If I2 > 50%, a random effects model was used to pool the results; otherwise, a fixed-effect model was applied  and its 95% confidence interval (CI) were used as the measuring standard to evaluate the strength of the association of PTC with the aggressive clinicopathologic features using fixed-effect and random-effect models. We measured heterogeneity among individual effect estimates, and we reported the P-value of the χsistency  and I2 i applied . To furt applied . The pub applied , 30. AllFrom the initial electronic search, we identified 763 articles, while from a manual search of the bibliographies and reference lists of these articles, we identified one additional article. After duplication removal, there were 250 records to screen for titles and abstracts. Following titles and abstract screening, 230 articles were excluded based on the inclusion and exclusion criteria. The full texts of the remaining 20 articles were checked for eligibility, and of these, eight were excluded –38 for vThe characteristics of studies included in the qualitative synthesis and meta-analysis are described in Egger’s test <0.05)   , and there was no heterogeneity among study results   . The results of the studies that were not included in our meta-analysis are shown in Among the 11 studies included, five studies were eligible for meta-analysis for the association between BMI and tumour size. A random-effect model was applied to analyse the data, and the overall OR for the overweight patients was 1.25 (95% CI: 0.97-1.53), and there was statistically significant high heterogeneity among study results (Ip=0.952) . The res2 = 59.5%, p=0.011)  . The results of the studies that were not included in our meta-analysis are described in Among the 11 studies included, seven studies were eligible for meta-analysis for the association between BMI and ETE. A random-effect model was applied to analyse the data and the overall OR for overweight patients was 1.26 (95% CI: 1.09-1.44), and there was statistically significant heterogeneity among study results (Ip=0.397) . The res2 = 47.1%, p=0.049)  . The results of the studies that were not included in our meta-analysis are presented in Among the 11 studies included, eight studies were eligible for meta-analysis for the association between BMI and multifocality. A fixed-effect model was applied to analyse the data, and the overall OR for overweight patients was 1.17 (95% CI: 1.10-1.24), and there was marginal statistical heterogeneity among study results (Ip=0.492) . The res2 = 58.6%, p=0.018)  . The results of the studies that were not included in our meta-analysis are shown in Among the 11 studies included, six studies were eligible for meta-analysis for the association between BMI and LN Metastasis. A random-effect model was applied to analyse the data, and the overall OR for overweight patients was 1.04 (95% CI: 0.90-1.17), and there was statistically significant heterogeneity among study results (Ip=0.059) . The resThe present study is the first systematic review and meta-analysis comprehensively evaluating the association of BMI with aggressive clinicopathologic features in patients with PTC. Eleven retrospective cohort studies involving 26,196 participants were included. Our findings showed that elevated BMI was significantly associated with ETE in both overweight  and obesity group  as well as with multifocality in both overweight  and obese patients . Likewise, obesity was significantly associated with increased tumour size  and with LN metastasis , whereas being overweight was significantly associated with a more advanced TNM stage .Thyroid cancer incidence has been rising, along with obesity rates over the past decade . Thyroidin vitro results in increased cell proliferation, and in thyroid cancer cells, IGF receptors were found to be overexpressed compared to normal thyroid cells , which has been linked to thyroid cancer through insulin-like growth factor-1 (IGF-1) and thyroid-stimulating hormone (TSH) signalling, inflammation, or oxidative stress . Hyperinid cells .Adiponectin, a collagen-like polypeptide, is a hormone involved in various metabolic processes, including insulin-sensitizing and anti-inflammatory processes. It is secreted from adipose tissue and acts through two transmembrane receptors (Adipo R1 and Adipo R2). It also acts as an anti-tumour-agent by inhibiting cell proliferation and angiogenesis and increasing apoptosis. In many studies, adiponectin was found to be inversely related to BMI. Although, a previous study revealed lower circulating adiponectin levels in any form of thyroid cancer , yet, a Another possible mechanism that has been proposed to explain the association between obesity and thyroid cancer is the elevated TSH that has been associated with both obesity and advanced tumour stage . TSH levvia the PI3K/AKT pathway through the membrane receptor Ob-R  other than the WHO-recommended BMI categories for relative weight classification as well as studies that grouped PTC with other histologic types of thyroid cancer, which could limit our findings. Of note, previous studies that examined the association between BMI and aggressiveness of differentiated thyroid cancer (DTC) found no significant associations between BMI with any aggressive clinicopathologic features of thyroid cancer , 65, 66.To the best of our knowledge, this is the first systematic review and meta-analysis to identify whether adiposity is related to aggressive clinicopathologic features in patients with PTC. Higher BMI is found to be associated with ETE and multifocality. Obesity was associated with tumor size and LN metastasis, whereas being overweight was associated with the advanced TNM stage. More studies with a larger number of participants are needed to examine and further clarify the association of increased BMI, advanced TNM stage, and LN metastasis. Given the current obesity epidemic and the solid evidence that adipose tissue supports both tumorigenesis and metastasis , 68, a sThe original contributions presented in the study are included in the article/AE carried out literature searches, appraised the articles, summarized the results, prepared the tables, wrote the manuscript and interpreted the results. KG supervised the process, carried out literature searches, appraised the articles, performed the meta-analysis, prepared the tables, wrote the manuscript and interpreted the results. IM interpreted the results. PE appraised the articles, summarized the results, prepared the tables, wrote the manuscript and interpreted the results. PP conceived the theme, supervised the process, wrote the manuscript and interpreted the results. All authors contributed to the article and approved the submitted version.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Due to clinical and methodological diversity, clinical studies included in meta-analyses often differ in ways that lead to differences in treatment effects across studies. Meta-regression analysis is generally recommended to explore associations between study-level characteristics and treatment effect, however, three key pitfalls of meta-regression may lead to invalid conclusions. Our aims were to determine the frequency of these three pitfalls of meta-regression analyses, examine characteristics associated with the occurrence of these pitfalls, and explore changes between 2002 and 2012.A meta-epidemiological study of studies including aggregate data meta-regression analysis in the years 2002 and 2012. We assessed the prevalence of meta-regression analyses with at least 1 of 3 pitfalls: ecological fallacy, overfitting, and inappropriate methods to regress treatment effects against the risk of the analysed outcome. We used logistic regression to investigate study characteristics associated with pitfalls and examined differences between 2002 and 2012.Our search yielded 580 studies with meta-analyses, of which 81 included meta-regression analyses with aggregated data. 57 meta-regression analyses were found to contain at least one pitfall (70%): 53 were susceptible to ecological fallacy (65%), 14 had a risk of overfitting (17%), and 5 inappropriately regressed treatment effects against the risk of the analysed outcome (6%). We found no difference in the prevalence of meta-regression analyses with methodological pitfalls between 2002 and 2012, nor any study-level characteristic that was clearly associated with the occurrence of any of the pitfalls.The majority of meta-regression analyses based on aggregate data contain methodological pitfalls that may result in misleading findings.The online version contains supplementary material available at 10.1186/s12874-021-01310-0. Due to clinical and methodological diversity, clinical studies included in meta-analyses often differ in ways that lead to differences in treatment effects across studies . Thus, aFirst, in meta-regression on aggregate data, associations between average patient characteristics and the pooled treatment effect do not necessarily reflect true associations between the individual patient-level characteristics and treatment effect , 5. The meta-analysis OR systematic[sb], as used in previous meta-epidemiological studies [We searched Medline through PubMed using the following search terms:  studies , 15 to s studies , 16 had Eligible were all aggregate-level meta-analyses published in 2002 and 2012 in which at least one study was a randomised or quasi-randomised controlled trial, which reported on a meta-regression analysis  examininA standardised and piloted data extraction form accompanied by a codebook was used for data extraction. We determined whether meta-regression analyses included at least one of three previously described pitfalls of meta-regression: (1) ecological fallacy, , 5 (2) aAdditional data extraction, done by a single author (MG), involved extraction of the following characteristics: the name of the journal and journal characteristics , affiliation of the authors with industry, affiliation of the authors with an institute with statistical expertise (mainly biostatistics or epidemiology department), the number of studies included in the review, design of the studies included in the review and in the meta-regression analysis, whether the outcome was continuous or binary, the medical field of speciality, and the category of therapy . For all reports, we used the impact factor in the year 2007 (midpoint between 2002 and 2012). Our operationalization of these variables is reported in Additional file Our sample of 2,404 records per entry year  was based on a pilot study of a random sample of 50 records from 2012, in which we identified 20 (40%) meta-analyses, 3 of which included a meta-regression analysis . Assuming that the yield of meta-regression analyses in 2002 would be about one third of the yield in 2012, we estimated that the yield of meta-regression analyses in the total random sample of 4,808 citations would be 192 . Our target sample size of 192 would yield 95% confidence intervals ranging from 14 to 26% if the prevalence of a potential pitfall was 20%, and ranging from 43 to 57% if the prevalence of a potential pitfall was 50%.The primary outcome was the proportion of meta-analyses with at least one meta-regression analysis that was potentially influenced by one of the 3 assessed methodological pitfalls in our random samples for the years 2002 and 2012. Secondary outcomes were the proportion of meta-analyses with at least one meta-regression analysis with ecological fallacy; the proportion of meta-analyses with at least one meta-regression analysis with a high risk of overfitting; the proportion of meta-analyses with at least one meta-regression analysis with inappropriate methods to regress treatment effects against the risk of the analysed outcome observed in included studies; and the proportion of meta-analyses with at least one meta-regression analysis that recognized the limitations when these pitfalls were present. We used Firth’s logistic regression to investigate the association between study characteristics and the occurrence of pitfalls, using one predictor at a time. Firth’s logistic regression model uses a penalized maximum likelihood that produces more accurate inference in small samples than the standard maximum likelihood based logistic regression estimator, and achieves convergence in the presence of separation with better coverage probability , 18. OddOur search of 2002 and 2012 reports resulted in 7,000 citations for 2002 and 20,169 citations for 2012  were affected by at least one of three key pitfalls of meta-regression analyses addressed in our study—ecological fallacy, overfitting, or inappropriate methods to regress treatment effects against the risk of the analysed outcome. This suggests that about one out of ten meta-analyses are potentially flawed and may report invalid findings. In only two of the 57 meta-regression analyses with pitfalls, did authors explicitly acknowledge the limitations of their findings derived from meta-regression analysis. We found little evidence for associations between inappropriate meta-regression and characteristics of meta-analyses. There was inconclusive evidence with regard to the association between journal, author, and characteristics of meta-analyses and the odds of pitfalls of meta-regression. The negative association between the prevalence of pitfalls in meta-regression analyses and authors affiliated with biostatistics or epidemiology department , although imprecise, is noteworthy. In addition, authors with such affiliation were less frequent in analyses published in 2012 (37%) than in 2002 (55%). This is striking and may reflect an increase in the availability of software for non-statisticians to conduct meta-regression analysis. This underlines the importance of a collaboration with a statistician experienced in evidence synthesis when conducting meta-regression analyses . In addiOur meta-epidemiological study has several strengths. First, we used a systematic and broad search strategy using a validated filter to find systematic reviews and meta-analyses. We refrained from using filters specific to meta-regression, as many meta-regression analyses were not mentioned in titles and abstracts of eligible meta-analyses. Second, our use of two random samples of meta-analyses from 2002 and 2012 suggests generalizability of our findings to published meta-analyses with meta-regression indexed in PubMed. Third, we conducted a sample size consideration prior to our screening, which was recently proposed by Giraudeau et al.  who alsoOur study has limitations. First, the number of meta-analyses with meta-regression analysis identified in our random samples was only 81, and therefore at the lower end of what we had expected based on our sample size consideration, which in turn has limited our statistical power to detect associations between pre-specified study characteristics and pitfalls. Second, our findings on overfitting may underestimate the true frequency of this pitfall as the threshold of less than 5 trials per covariate was more conservative than currently suggested in the Cochrane Handbook . Third, To our knowledge, no prior meta-epidemiological study has quantified the number of meta-regression analyses affected by these pitfalls. The problem of ecological fallacy in meta-regression analysis is well known , 23, 24.The minimum number of trials per covariate in meta-regression analyses required to minimize the risk of overfitting is unknown. The Cochrane Handbook suggest a minimum of 10 studies per examined covariate,  but a reMeta-regression is often used to determine whether treatment effects are associated with the underlying baseline or population risk using the event rate observed in the control group as a surrogate for baseline risk , 9. ThisMuch like the decision to conduct a fixed-effect or random-effects meta-analysis, the decision to conduct a meta-regression analysis should not be based on heterogeneity assessments using, for instance, a chi-squared test, Cochrane Q, or I-squared . InsteadResults of the majority of meta-regression analyses based on aggregate data may be misleading. A considerable body of methodological literature and recommendations appear to have had little impact on the use of meta-regression in the medical literature. Authors, editors and peer reviewers need to become more aware of the methodological pitfalls of meta-regression analyses.Additional file 1


ESR1 rs9340799 polymorphism has been frequently investigated with regard to its association with breast cancer (BC) susceptibility, but the findings have been inconclusive. In this work, we aimed to address the inconsistencies in study findings by performing a systematic review and meta-analysis. Eligible studies were identified from the Web of Science, PubMed, Scopus, China National Knowledge Infrastructure, VIP and Wanfang databases based on the predefined inclusion and exclusion criteria. The pooled odds ratio (OR) was then calculated under five genetic models: homozygous (GG vs. AA), heterozygous (AG vs. AA), dominant (AG + GG vs. AA), recessive (GG vs. AA + AG) and allele (G vs. A). Combined results from 23 studies involving 34,721 subjects indicated a lack of significant association between the polymorphism and BC susceptibility . Subgroup analyses by ethnicity, menopausal status and study quality also revealed no statistically significant association (P > 0.05). In conclusion, our results showed that the ESR1 rs9340799 polymorphism was not associated with BC susceptibility, suggesting its limited potential as a genetic marker for BC.The  Like other malignancies, risk factors of BC are primarily genetic predisposition and environmental influences3. It has been reported that genetic background or familial history accounts for ~ 20–25% of overall BC incidence4. Among the ~ 80 genetic loci known to be associated with susceptibility to BC, the BRCA1 and BRCA2 loci carry the highest risk6. Together with other high- and medium-penetrance germline mutations located at the loci of TP53, PTEN, ATM, BRIP1, CHEK2 and PALB2, they made up ~ 15–20% of the genetic risk of BC9. Common low‑penetrance genetic polymorphisms account for the remaining risk for BC susceptibility11. On the other hand, environmental and lifestyle risk factors for BC include the consumption of oral contraceptive, cigarette smoking, alcohol consumption, breastfeeding and delayed age at first childbirth14.According to the World Health Organization statistics, breast cancer (BC) is the most common malignant tumor type, as well as a leading cause of mortality in the female population16. Mounting evidence from population-based studies has corroborated the association of endogenous and exogenous circulating estrogen in BC etiology and the increased risk of BC in premenopausal women17. The physiological receptors for estrogen are estrogen receptors (ER), which function to mediate the effect of estrogen on breast cells. Binding of estrogen to ER promotes the growth and differentiation of the normal breast cells and can lead to breast carcinogenesis18. There are two ER isoforms, i.e., ERα and ERβ. These two ER isoforms are respectively encoded by two distinct genes, ESR1 and ESR219. ERα has a higher level of expression in the breast tissue between these two isoforms, hence it is frequently implicated in BC development20.Among these risk factors, it has been specifically pointed out that estrogen can act as a carcinogen, not only by causing chromosome segregation errors as well as structural chromosomal alterations, but also by stimulating the uncontrolled proliferation of mutated breast cellsESR1, which is highly polymorphic. Among the many polymorphisms in ESR1, the two best-studied ones are rs2234693  and rs9340799  polymorphisms. Both polymorphisms are located in intron 1, respectively at 1,397 bp and 351 bp upstream of exon 2 of the gene, and have been associated with several female cancers, including BC and endometrial cancer23. However, the association of the two polymorphisms with BC susceptibility has been described with conflicting results in many studies27. To sort out this inconsistency, a meta-analysis on rs2234693 was performed in 2018, and showed that the polymorphism was significantly associated with a decreased BC susceptibility28. As for rs9340799, a meta-analysis based on seven previous studies was reported by Zhang et al. in 2015, and found no significant association between the polymorphism and BC susceptibility under all three genetic models examined, even when the data were stratified into subgroups according to the ethnicity and source of controls29. In this current work, we attempted to perform an updated meta-analysis on the relationship between ESR1 rs9340799 polymorphism and BC susceptibility, by including a large number of additional studies that have been left out by Zhang et al. or have only been published after 2015.The focus of this meta-analysis is the ERα-encoding gene, 30 were excluded. In addition, two articles were found to contain overlapping data31, and the one with the smaller sample size was excluded31. Ultimately, 22 articles comprising 23 studies were included for the quantitative synthesis of data47.The study selection process is depicted in Fig. 41, and four other studies included only postmenopausal women40. The remaining studies either did not mention the menopausal status or did not perform separate analyses for pre- and postmenopausal women. In terms of ethnicity, nine studies were conducted on Asians44, nine on Caucasian47, three on other ethnicities46, and two on mixed ethnicities42. All studies were case–control in design. Fifteen (15) of the studies were considered as having high quality, whereas eight had low quality . Among the included studies, eight (from seven articles) reported data for pre- and postmenopausal women separatelyESR1 rs9340799 polymorphism and BC susceptibility . The random-effects model was used in the above analyses as significant heterogeneity was present in all genetic models. The forest plots of the associations are presented in Fig. The meta-analysis results are shown in Table Subgroup analyses were performed based on the ethnicity (Asian vs. Caucasian) and menopausal status (premenopause vs. postmenopause) of the study subjects, as well as the quality of the studies . No statistical significant association was observed for all subgroups under all genetic models , premenopause (I2 = 0.0%), postmenopause (I2 = 0.0%) and low quality (I2 = 19.8%) subgroups. A similar observation was observed for the recessive model . In heterozygous model, the Caucasian (I2 = 18.8%) and high quality (I2 = 34.9%) subgroups showed low heterogeneity, whereas in allele model, low heterogeneity was noted in premenopause (I2 = 24.5%), postmenopause (I2 = 47.5%) and low quality subgroups (I2 = 46.7%). All subgroups in the dominant model showed high heterogeneity (I2 > 50%).Although significant heterogeneity was observed in the overall analysis, several subgroups were found to have low heterogeneity based on the No evidence of asymmetry was detected in the funnel plots of all genetic models Fig. , indicatESR1, which is mapped to chromosomal locus 6q25.1 and contains eight exons. It has been documented that the human ESR1 gene contains at least nine promoters, whereby each promoter harbors multiple transcription factors-binding sites48. The ERα protein possesses DNA- and ligand-binding domains which are highly conserved49. It is depicted that ERα can mediate the effect of estrogen via several molecular pathways. Among these, the classical pathway is the best-known. In this direct pathway, unliganded ERα forms a cytosolic complex with Hsp90. Upon estrogen binding to the ligand-binding domains of ERα, the ERα-Hsp90 complex dissociates. Subsequently, ERα dimerizes and translocates to the nucleus. Following that, the DNA-binding domains of ERα, consisting of two functionally distinct zinc finger motifs, bind to a characteristic stretch of DNA sequence named the estrogen response elements in the promoters of the target genes to influence the process of transcription50.ERα, a member of the nuclear receptor superfamily, is encoded by a ~ 300 kb gene, 51. The nongenomic pathway typically involves a small plasma membrane population- and cytoplasm-based ERα52, which interacts with signaling proteins such as Src, mitogen-activated protein kinase (MAPK) and phosphoinositide 3-kinase. These signaling molecules can activate the phosphorylation of ERα and its coregulators54. This subsequently triggers signaling cascades via second messengers (SM), and eventually, it enhances nuclear ERα signaling without involving gene regulation. The last ER pathway is the ligand-independent pathway. In this case, ERs can become activated via crosstalk with other signaling pathways, e.g. the insulin-like growth factor-1 receptor and the epidermal growth factor receptor pathways55. In these instances, ERs are activated by phosphorylation to form dimers, to bind DNA, and regulate the expression of genes.Meanwhile, the tethered pathway entails protein–protein interaction or heterodimerization of ERα with other transcription factors such as AP1or NF-kB after ligand activation. This results in the indirect binding of DNA by ERα, contributing to the regulation of target genes including insulin-like growth factor 1, cathepsin D, progesterone receptor, transforming growth factor α, pS2, retinoic acid receptor α1, c-myc, etc., which are essential for cell proliferation and survival54. ER has been used as a molecular classifier for breast tumors, whereby BC can be graded as ER-positive and ER-negative. A large proportion (~ 75%) of BC are known to be ER-positive56. ERα-positive cases are often associated with more optimistic prognosis as they generally respond more positively to endocrine therapies, and are also sensitive to CDK4/6 inhibitors57. In contrast, ERα-negative BC is generally regarded as aggressive and metastatic malignancies58.Notwithstanding, all models of ERα signaling pathways point to the vital role of ERα in the proliferation and survival of breast epithelial cells, as well as mammary tumorigenesis59. For this reason, many genetic association studies have investigated the relationship between ESR1 polymorphisms and BC susceptibility. These polymorphisms include, but not limited to, rs9340799, rs3020364, rs9322335, rs2234693, rs1801132, rs2046210, rs3020314, rs1514348, rs3020314, rs1514348, rs1514348 and rs302031463.Given the important role of ERα in BC, its level and structure need to be tightly regulated to ensure an optimal functionality. The level and structure of a protein are known to be influenced by, among others, genetic polymorphismsESR1. This is because the rs9340799 polymorphism has been widely studied and conflicting results have been frequently obtained, and no recent meta-analysis has been carried out to address the inconsistencies in the study findings. For instance, while Wang et al. reported that the GG genotype of the polymorphism was associated with a reduced susceptibility to BC, Sierra‑Martínez et al. reported that the same genotype was associated with an increased susceptibility to BC45. Besides, Sakoda et al. did not find any significant association between the polymorphism and BC susceptibility34. The difference in the study findings could be attributed to the variations in allele frequency across different studies. These variations are particularly relevant in populations consisting of different ethnicities, as interethnic differences in allele frequencies have long been known65. Taking the examples above, while Wang et al.36 noted in a Caucasian population that the minor allele frequency (MAF) of the polymorphism was 0.369, Sakoda et al.34 found that the MAF was merely 0.192 in an Asian population. These variations can account for differences in gene expression and therefore, disease susceptibility67. It is thus important to take into account the population variations in the allele frequency when attempting to identify a genetic biomarker for early identification of a disease68. For this reason, heterogeneity tests and subgroup analysis by ethnicity need to be performed when pooling the results from different studies together, as were done in our meta-analysis.Among these many polymorphisms, we have chosen to focus on rs9340799, an intronic polymorphism located just upstream of exon 2 of ESR1 may influence the cancer susceptibility by (i) being in linkage disequilibrium with another functional polymorphism in the same locus; (ii) influencing the expression of other genes through alterations to their transcriptional activity or mRNA stability; (iii) containing regulatory sequences which can impact gene expression via transcriptional regulation69. For these reasons, in this meta-analysis, we attempted to precisely re-examine the relationship between the ESR1 rs9340799 polymorphism and the susceptibility to BC. In doing so, we included 23 case–control studies from 22 systematically selected published articles. We performed the meta-analysis under five different genetic models, namely the homozygous, heterozygous, dominant, recessive, and allele models. Importantly, our analyses with all five genetic models failed to detect any significant association between the rs9340799 polymorphism and BC susceptibility. Under each genetic model, we further stratified our analysis based on the following subgroups: (i) ethnicity (Asian vs. Caucasian), (ii) menopausal status (premenopause vs. postmenopause), and (iii) study quality . Again, none of these subgroups showed any significant association. Notably, our finding was in agreement with that of the Zhang et al. even though we have included more studies (N = 23 vs. N = 7)29.It is noteworthy that most of these studies have centered on genetic association rather than deciphering the exact biological mechanisms. Nonetheless, it has been postulated that intronic polymorphisms such as the rs9340799 polymorphism of ESR1. The analyses of more polymorphisms of ESR1 in future, either individually or in tandem, may further reveal the synergistic effects of such polymorphisms in influencing BC susceptibility70.The major strength of our study is that we have analyzed data from a large population of meticulously selected studies; therefore, this study has strong statistical power. Besides, the chosen exposure, i.e., the rs9340799 polymorphism, is a discrete and well-defined parameter that can be genotyped with high precision using the available technologies. This allows a fair comparison to be made among independent studies, contributing to more consistent inter-laboratory or inter-study comparison. On the other hand, the major limitation of this study is that gene–gene or gene-environment interactions were not measured as most of the included studies did not report this information. Furthermore, our meta-analysis has so far focused on one polymorphism from ESR1 and the susceptibility to BC, despite the different genetic models considered. Each genetic model was further divided into subgroups based on ethnicity, study quality and menopausal status, but similarly, no statistically significant association was observed. Nevertheless, our conclusion warrants further studies, given that the ESR1 harbors many polymorphisms that await detailed investigation.In conclusion, our overall results revealed no significant association between the rs9340799 polymorphism of ESR1 rs9340799 polymorphism and BC susceptibility, and (ii) reported the genotype and allele frequencies or contained necessary data to obtain the information. Studies were excluded if (i) they were not original research papers (e.g. review articles or commentaries), and (ii) the investigations were not performed on human subjects. The reference lists of the eligible studies were also manually screened to identify additional relevant studies. When overlapping data were found, we included only the study with the largest sample size. The study protocol was pre-registered with PROSPERO (registration number: CRD42021231912).A comprehensive literature search was performed in the Web of Science (WoS), PubMed, Scopus, China National Knowledge Infrastructure (CNKI), VIP and Wanfang databases up to January 21st, 2021, without language restriction. The following search terms were used: (ESR1 OR estrogen receptor) AND (XbaI OR rs9340799) AND (polymorphism or variant) AND (breast cancer OR breast neoplasm). Studies were selected if they fulfilled the following inclusion criteria: (i) were case–control and/or cohort studies which have investigated the association between 71. Studies rated ≥ 6 stars were considered high quality.Three investigators independently extracted the following data from the included studies: name of the first author, publication year, location, ethnic group, sample size, genotype and allele frequencies, menopausal status, genotyping method, blinding status, genotyping success rate, and sources of controls. Discrepancies were resolved through discussion until a mutual agreement was reached. The P-values of the Hardy–Weinberg equilibrium (HWE) among the control group was calculated using a goodness-of-fit test. The Modified Newcastle–Ottawa Scale for Case–Control Studies of Genetic Association was used to assess the quality of the included studiesESR1 rs9340799 polymorphism and BC susceptibility was evaluated using the odds ratio (OR) for various genetic models, i.e. homozygous (GG vs. AA), heterozygous (AG vs. AA), dominant (AG + GG vs. AA), recessive (GG vs. AA + AG) and allele (G vs. A). A forest plot was also generated to graphically represent the findings. A fixed-effect model was used if the heterogeneity among the studies was low . On the other hand, when heterogeneity was significant, a random-effects model was used. Sensitivity analysis was performed using the leave-one-out method for evaluating the robustness of the findings. Subgroup analyses were performed according to ethnicity (Asian vs. Caucasian), study quality , and menopausal status (premenopause vs. postmenopause). In most included studies, the ethnicity was explicitly stated, although the standards of classification  was not known. However, when such information was not available, the populations were classified into different ethnicities based on the major ethnic group of the countries in which the subjects were recruited. Publication bias was evaluated using the Begg’s and the Egger’s tests, and through visual inspection of the funnel plot for asymmetry. For all analyses, the result was considered to be statistically significant when P < 0.05, unless otherwise stated.STATA version 16.0  was used for the quantitative synthesis of the data. The association between Supplementary Information.


No significantly increased risk due to obesity was found for women. A 49% increased CRC risk for obesity was found for subjects with an MLH1 mutation . These results confirm the different effects of sex on obesity and CRC risk and also support the public measures to reduce overweight in people with LS, particularly for men.There appears to be a sex-specific association between obesity and colorectal neoplasia in patients with Lynch Syndrome (LS). We meta-analyzed studies reporting on obesity and colorectal cancer (CRC) risk in LS patients to test whether obese subjects were at increased risk of cancer compared to those of normal weight. We explored also a possible sex-specific relationship between adiposity and CRC risk among patients with LS. The summary relative risk (SRR) and 95% confidence intervals (CI) were calculated through random effect models. We investigated the causes of between-study heterogeneity and assessed the presence of publication bias. We were able to retrieve suitable data from four independent studies. We found a twofold risk of CRC in obese men compared to nonobese men (SRR = 2.09; 95%CI: 1.23–3.55, I MLH1, MSH2, MSH6, or PMS2) or a germline deletion in an epithelial cell adhesion molecule (EPCAM), which leads to epigenetic inactivation of MSH2 [Colorectal cancer (CRC) accounts for approximately 10% of all annually diagnosed cancers and cancer-related deaths worldwide . LS runs of MSH2 ,5. Depen of MSH2 . The bio of MSH2 . In the  of MSH2 . However of MSH2 ,12. In p of MSH2 ,14, we pThis literature review and meta-analysis was designed, conducted, and described according to the MOOSE guidelines for systematic reviews and meta-analyses of observational studies ; the stuWe carried out a meta-analysis based on published evidence to investigate whether sex and MMR genes may modify the effect played by obesity on the development of CRC in patients with LS. The primary inclusion criteria identified for potentially eligible studies for this meta-analysis were the following: (i) the studies should include subjects with LS or subjects who met the Amsterdam/Bethesda (revised) criteria for LS ; CRC ris2 statistic, which ranges from 0% to 100%, was provided to quantify the percentage of total variation across studies that is attributable to heterogeneity rather than chance. A threshold of I2 below 50% was considered an acceptable level of between-study heterogeneity [As CRC is a rare event in the population, all estimates of risk  for obese versus non-obese subjects were considered a good approximation of relative risk (RR). Every measure of association and its 95% CI were log-transformed and the corresponding standard error was calculated by using the formula proposed by Greenland . Summaryogeneity . A possiogeneity . A sensiogeneity , which iAccording to the “Population-Item-Comparison-Outcome” (PICO) framework, the population of interest was exclusively composed of people with LS or subjects who met the Amsterdam/Bethesda (revised) criteria. The exposure of interest was the sex-specific relationship between adiposity and colorectal neoplasia. The literature review was conducted in Medline and EMBASE for papers published up to 31 December 2019, using the following search string: (“lynch” OR hereditary non-polyposis colon cancer” OR “HNPCC”) AND (“weight” OR “obesity” OR “BMI” OR “adipose” OR “adiposity”) AND  AND (“cancer” OR “tumor”). The controlled vocabulary and keywords included in the search string were designed to ensure that all published evidence regarding weight/BMI and CRS in subjects with LS would be covered. After removing duplicates entries, an initial screening based on title and abstract was made independently by three researchers , and papers were discarded when there was consensus among the panel members. Papers that were instead deemed potentially suitable by at least one researcher were read in full by all panel members, who independently verified that all inclusion criteria were met. Any disagreements were resolved by consensus. Bibliography of relevant studies was also checked to further improve the search and confirm that all information was recruited. No time or language restriction was applied. Three independent authors  who selected studies and extracted relevant data conducted a comprehensive literature search.The literature search produced 51 items. Eighteen were excluded by the consensus panel. depending on the title and the abstract of a research paper, . We readOne case-control study  was not 2). For Win et al. [MLH1 and MSH2 genes. All studies with a prospective study design expressed the association between obesity and CRC in terms of adjusted HR (95% CI), while the case-control study reported the adjusted OR (95% CI). Three of the selected studies ,25,26 ar2 = 33%), with no indication of publication bias (p = 0.13). No significant difference was found between obese versus nonobese women .All the studies assessed the risk for CRC, except for Botma et al.  who evalMLH1 and MSH2, for increasing the value of BMI by 5 kg/m2 , with no indication of heterogeneity, I2 = 0%. No association was found for MSH2: summary RR: 1.15 (95%CI: 0.94–1.41) every increase of 5 kg/m2, with no indication of heterogeneity, I2 = 0%. The quality of the studies was generally very high  appears to be a sex-limited trait with rather low penetrance in men [PMS2 variants had no increased risk of CRC, irrespective of sex  [Very recent findings from a prospective cohort of 6350 carriers of pathogenic mutations in MMR genes have now imposed a revision of the management guidelines, considering different gene and sex-specific risks . The lifypectomy . The rise in men . Lastly,er ages) . MLH1 carriers. Very recently, data suggested possible different pathways of CRC development in MLH1 vs. MSH2 carriers [MSH2-associated CRCs have a higher frequency of somatic APC mutations compared with MLH1-associated CRCs, a significantly higher frequency of CTNNB1 mutations has been observed in MLH1-associated CRCs compared with MSH2-associated ones [CTNNB1-negative CRCs and with every 5 kg of body fat (measured by BMI), the risk for CTNNB1-negative colorectal cancer increases by about one-third [In addition to nonmodifiable factors, such as genes and sex, we found that adiposity may also significantly increase the risk of CRC in men with a genetic predisposition, thus better depicting the concept of “incomplete penetrance”. This is a term often criticized as reflecting the lack of knowledge of genetic and environmental factors that may interact with the genotype to determine the ultimate phenotype of the individual. It is even more intriguing to find that the increased risk due to adiposity in subjects with LS seems to be limited to path_carriers . In partted ones . Obesityne-third . All theCOX-1, and a reduced total mitosis in the crypt in the colorectal mucosa [The guidelines committee of NICE published draft guidance advising physicians that people with LS should be offered daily aspirin to reduce their risk of CRC . The CaPl mucosa . These results might be achievable by promoting weight loss by lifestyle changes  . A feasiPMS2 and path_MSH6 carriers, where, precisely because of the low penetrance of these two genes, adiposity could have been a decisive modifier. Finally, Botma et al. [The main strengths of our meta-analysis are the fullness of the literature search, despite the number of the included studies on this specific topic, and the calculation of summary CRC risk estimates in both obese men and women with LS. Despite the good methodological quality of most included studies, there is a substantial variability in terms of study design, methods used to define exposure definition, and statistical methods, potentially affecting the study results. This resulted in a considerable between-estimates heterogeneity, thus highlighting the necessity to standardize methods to get comparable risk estimates and reach more robust conclusions. Another limitation of all the included studies was that height and weight  were self-reported. Since it is well known that height and weight are often overestimated and underestimated, respectively, absolute rates of obesity are frequently underestimated . Self-rea et al.  analyzeda et al. .Despite the abovementioned limitations, because the available evidence about the effects of weight loss on CRC risk is limited, especially in subjects with LS, along with the difficulty to perform large randomized clinical trials in patients with rare conditions such as LS, we think our meta-analysis takes on extra significance and supports future research. In the meantime, public measures to reduce the development of obesity and to enable those who are obese to lose weight (at least 20% of their body weight)  are mand


Although there are reports that artificial sweeteners (AS) are safe, the relationship between artificial sweeteners and cancer remains controversial. The purpose of the study is to evaluate whether the consumption of artificial sweeteners is associated with the risk of cancers. We conducted a comprehensive search of multiple databases, including MEDLINE, EMBASE, Web of Science, and Cochrane Library. We found all the literature that studied the relationship between artificial sweeteners and cancer. Ten case–control studies were included in the meta‐analysis. Our findings indicated that the consumption of artificial sweeteners was not associated with an increase in cancer when all types of cancers are analyzed comprehensively . Interestingly, the use of artificial sweeteners is inversely related to urinary system cancer risk when analyzing women individually . Our meta‐analysis found that these is no correlation between artificial sweeteners and occurrence of cancer except urinary system cancer in women. Considering some limitations found in this study, additional data from large clinical trials are needed. Although there are reports that artificial sweeteners (AS) are safe, the relationship between artificial sweeteners and cancer remains controversial. The purpose of the study is to evaluate whether the consumption of artificial sweeteners is associated with the risk of cancers. The following three exclusion criteria were applied: (1) We excluded studies that did not provide initial data, animal studies, in vitro studies, reviews, letters, personal opinions, book chapters, and conference abstracts; (2) only show the consumption of drinks containing artificial sweeteners; and (3) studies that full paper copy were not available. Two investigators independently reviewed the literature, extracted all potentially eligible studies, and resolved uncertainty and disagreement by discussion  first author, nationality, publication year, numbers, and age of case subjects and the control group; and (2) artificial sweetener type, the number of people exposed to artificial sweeteners in case and control groups, and the type of cancer. Any differences related to the data extraction were resolved by rechecking the full text of the study or by discussion. When study data were ambiguous or data were not reported in a form that could be used for formal comparison, we contacted the corresponding and first author of the original publication via email.2.4The methodological quality for the selected literature was evaluated independently by two investigators according to the Newcastle–Ottawa scale (NOS) Stang, . The NOS2.5p < .05). We calculated the odds ratios (ORs) and their respective 95% confidence intervals for case–control studies. The random‐effect model was used to compute OR and 95% confidence intervals (CIs) and to assess the differences of artificial sweetener exposure between the case group and control group. Cochran's Q statistic and the I2 statistic were used to assess the statistical heterogeneity in the meta‐analysis , a fixed‐effect model meta‐analysis was performed; if the data were heterogeneous (p ≤ .05), a random‐effects model meta‐analysis was performed. In the Q test, p < .05 was considered significant for heterogeneity, and the I2 value was used to evaluate the degree of heterogeneity. I2 values of 25%, 50%, and 75% indicate low, moderate, and high heterogeneity, respectively . Normalization test showed that the data included in this article do not conform to the normal distribution  showed that the effect of publication bias was considered slight  ; however, the use of artificial sweeteners is inversely related to the risk of urinary system cancer in women , which is different from previous research. The reason for this result may be that the sample size of women is small or that the number of women in the case group is less than one‐third of the number of men or the number of women in the control group is less than half of the number of women. In addition, the short average use time of artificial sweeteners in the population included in the study may also be a reason. From the research data of urinary system cancer, it can be found that among the population included in the study, there are fewer women than men, which reflects from the side that the lower prevalence of female urinary system cancer. A researcher Morrison et al.  found thIn addition to observational research, there are people who have conducted genetic research. One study reported the interaction of aspartame and its metabolites with DNA in an in vitro model (Karikas et al., The relationship between the consumption of artificial sweeteners and cancer is a complex research topic, because the range of artificial sweeteners and cancer is very wide. Human data on artificial sweetener intake and cancer risk are scarce and largely have not been supportive of an association between artificial sweetener intake and cancer risk (Bosetti et al., Most of the previous research was on the relationship between artificial sweeteners and urinary system cancer. Toews et al.  analyzedCase–control studies have some inherent shortcomings, such as more research bias and confounding effects. Due to the small amount of literature on the relationship between artificial sweeteners and cancer, this meta‐analysis has certain limitations. Only 10 case–control studies with initial data were retrieved, the heterogeneity of result is high, and the reason may be the differences of subjects. In addition, this article used “yes” or “no” to determine whether artificial sweeteners have been consumed, without considering the consumption of amount and time of artificial sweeteners. Importantly, the scores of the INOS scale are not high. A previous researcher suggested that future research should evaluate the health effects of using artificial sweeteners with appropriate research time (Toews et al., 5In summary, there is no sufficient evidence to show whether the use of artificial sweeteners increases or decreases the risk of cancer. Considering some limitations found in this study, more data from large clinical trials are needed to affirm the relationship between artificial sweeteners and cancer.The authors declare that there are no conflicts of interest regarding the publication of this paper.WC, DL, LL, and PZ made the study design. LL and PZ conducted the study. LL, PZ, and YW analyzed the data and wrote the manuscript. LL, PZ, and YW attended the manuscript revision. All authors agreed with the final manuscript.Ethical Review: This study does not involve any human or animal testing. Ethics approval was not required for this research.Supplementary MaterialClick here for additional data file.


Survivin has been recently identified as a promising novel therapeutic target and prognostic marker in different types of cancer. Here we conducted a comprehensive meta-analysis to better clarify they the precise prognostic and diagnostic value of survivin in head and neck squamous cell carcinoma (HNSCC).Database of PubMed (Medline), Embase, and Web of Science were systematically searched for related published literature up to September 2020. Pooled hazards ratios (HR) and related 95% confidence intervals (CI) were used to estimate the association of survivin expression and survival outcomes in HNSCC patients.P < 0.001), and poorer disease-free survival (DFS)/ disease-specific survival (DSS) . Similar results were observed in subgroup analysis stratified by different cancer types, such as laryngeal squamous cell carcinoma (LSCC) , oral squamous cell carcinomas (OSCC) , nasopharyngeal carcinoma (NPC)  and HNSCC . Furthermore, ethnicity-stratified analysis indicated that survivin was significantly associated with poorer OS among both Asian and Non- Asian HNSCC patients .Twenty eight studies with 4891 patients were finally included in this meta-analysis, the pooled analysis indicated that the survivin expression was significantly correlated with poorer overall survival (OS) (HR, 2.02; 95% CI, 1.65–2.47, Our results suggested that survivin is predictors of worse prognosis in HNSCC patients. Hence, survivin is a potential therapeutic target for HNSCC.The online version contains supplementary material available at 10.1186/s12885-021-08170-3. Head and neck squamous cell carcinoma (HNSCC) is ranking as the sixth most prevalent cancer worldwide , they deSurvivin, an important member of the ‘inhibitor of apoptosis’ (IAP) family, is encoded by baculoviral inverted repeats (BIR) C5 gene . SurviviAlthough numerous researches have revealed the association between survivin and the prognosis of HNSCC, the results still remain controversial due to the variance in the sample size, study design, test approach and cut-off value. Therefore, it is essential to clarify the diagnostic and prognostic value of survivin in HNSCC based on the findings from the independently small sample size publications. Here, we searched PubMed (Medline), Embase, and Web of Science databases for relevant publications and undertaken a comprehensive meta-analysis to systematically assess the diagnostic and prognostic role of survivin in HNSCC.We searched for articles published between 2000 and 2020. Electronic searches for relevant retrieve studies were performed throughout databases, including PubMed (Medline), Web of Science and EMBASE databases in accordance with Dickersin et al. in September 2020 . The retp values and Kaplan–Meier curves [This meta-analysis was limited to studies about the association between HNSCC and survivin. The inclusion criteria of the meta-analysis were as follows: (1) all patients should be diagnosed with HNSCC; (2) survivin was evaluated in both samples of the HNSCC and normal controls; (3) studies revealed the association between survivin and survival of HNSCC; (4) sufficient statistical analysis was required, including hazard ratios (HR) and their related 95% confidence interval (95% CI)) for survival outcomes, if not we could calculate them by r curves  (5) the We inspected the duplicates, and removed the repeated papers first. Then, we perused the titles and abstracts of the papers carefully. At last full articles were selected to include the appropriate studies. Two researchers independently evaluated the literature against the inclusion and exclusion criteria (LQ Zhou and Y Hu). Any discrepancy in assessments was resolved by consulting an adjudicating with a third researcher (HJ Xiao). The researchers of the studies were contacted by e-mail to request data or additional information for meta-analytic calculations. The eligible studies for this meta-analysis were reviewed by two reviewers (LQ Zhou and Y Hu) independently. The Newcastle–Ottawa Quality Assessment Scale (NOS) was included to assess the methodological qualities of each publication, a star system (range from 0 to 9) was adopted to evaluate a including publications in three domains, comparability of study groups (2 stars), selection of participants (4 stars) and the outcome measurement (3 stars). A study with NOS ≥ 6 was seemed as a high-quality study , 13. Repp values and other published data following Parmer’s methods [2-based Q test and the I2 statistics [P > 0.10, I2 < 50%). Otherwise, the random-effects model was applied. Moreover, we performed subgroup analysis to explore the potential source of heterogeneity. Sensitivity analysis was conducted to investigate the influence of each individual study on the overall pooled results. We used the Begg’s and Egger’s tests to assess the potential publication bias. We conducted all statistical analyses by STATA statistical software version 15.0 (StataCorp Lp).The HR and the related 95% CI of survival outcomes were obtained directly from the primary publications or estimated by  methods . Statistatistics . The fixn = 431), reviews, abstracts and letters (n = 25) and the studies of irrelevant topics (n = 244), 78 studies were remained for further assessment by reading their full-text articles. A total of 50 studies did not provide specific data regarding HNSCC or survivin and therefore were excluded. Finally, a total 4891 HNSCC patients in 28 studies with publication years ranging from 2002 to 2019 were enrolled in the present meta-analysis.A total of 778 potential records were initially identified by searching the electronic databases Fig. . FollowiThe characteristics of the eligible studies were summarized in Table P < 0.1). Hence, the random-effects model was adopted and subgroup analysis was used to seek for the potential causes of heterogeneity. The results of these studies indicated expression of survivin were associated with poorer OS . Medium heterogeneity was noted  , low heterogeneity was noted between survivin expression and DSS .A total of twenty one studies in the present analysis examined the association between survivin and the OS of in HNSCC patients. The heterogeneity among the publications in our study was significant for Q test (P < 0.001) with low heterogeneity ; for HNSCC was 1.52  with medium heterogeneity ; for NPC was 2.53  with no significant heterogeneity  and for OSCC was 2.45  with no significant heterogeneity   with medium heterogeneity ; for Asian patients was 2.16  with no heterogeneity  . Therefore, “trim and fill” analysis was further utilized, and the pooled HR of 1.569  remained statistically significant (Fig. The publication bias was assessed by the Begg’s funnel plots and the Egger’s test in the present study. The results indicated the publication bias existed among the studies (Numerous studies have focused on the identification of new prognostic biomarkers that can be used for cancer monitoring and detection. An association between survivin expression and survival outcomes has been shown in HNSCC patients. The study by Fiedler et al. found that high survivin expression was significantly correlated to unfavorable DSS in HNSCC patients . Kim et The prognostic value of survivin was investigated in HNSCC and gathering evidences suggested that survivin was an independent prognostic marker in HNSCC , 47. EpiHowever, the present meta-analysis also had several limitations. First, we found that different paper qualities and sample sizes across the studies might cause bias in the meta-analysis. Second, the approaches used to evaluate the survivin expression were different, such as antibodies characters in IHC and instruments used in RT-PCR. Third, the cut-off value defining positive survivin expression varied among eligible studies Table . Forth, In conclusion, we searched the electronic databases and a total of 4891 patients in 28 studies were enrolled for meta-analysis, the results demonstrating that patients with survivin expression are more likely to have worse prognosis. Taken together, our meta-analysis results suggest that survivin gains a prognostic and diagnostic value for the HNSCC patients. However, more larger sample size studies are required to acquire a more representative and precise result.Additional file 1: Table S1. Characteristics of the antibodies used in IHC in the including studies.


Dietetics today occupy a significant place in the field of research, helping to discover cardiovascular benefits of healthy diets and consumption of organic foods such as fruits, vegetables, legumes, nuts, and whole grains. One of the components of vegetable-based diet is chili pepper (CP) which has been found to affect all-cause mortality.MEDLINE, EMBASE, Scopus, EBSCO, and Cochrane (Wiley) Central Register of Controlled Trials were searched from inception till January 9, 2020, identifying all relevant studies using keywords and truncations. Studies were included if (1) they were observational or randomized in nature (2) included patients consuming CP and (3) evaluated direct comparison between regular and rarely/never CP consumption.Our preliminary search yielded 6976 articles. Post exclusion and after full-text screening, four potential observational studies with a population of 570,762. Pooled analysis found reduced all-cause mortality in CP consumers compared to nonconsumers with a risk ratio (RR) of 0.75 . The RR for CVD, cancer related and CVA deaths were 0.74 , 0.77  and 0.76 , respectively.Statistically significant results of our analysis put forward a rationale indicating an association between lower risk of all-cause, cardiovascular and cancer related deaths and CP consumption. •Cardiovascular diseases have been observed to be a major cause of mortality.•Recent transition to vegetable-based diet including Chili Pepper is positively impacting health of cardiovascular patients.•Capsaicin affects TRPV1 Receptors and consequently weight-reduction. One of the components of vegetable-based diet is chili pepper (CP) which has been found to affect all-cause mortality . The che2AMSTAR-2) tool . The RR for CVD, cancer related and CVA deaths were 0.74 , 0.77  and 0.76 , respectively  met the inclusion criteria and thus included in the meta-analysis . The stuectively .Table 1C4This is the first meta-analysis carried out to assess the impact of CP consumption on all-cause, CVD and cancer related mortality. Our results show significant benefit from CP consumption in preventing such deaths as opposed to rare or no CP consumption. The lack of data on mode, quantity and frequency of CP consumption leads to non-standardization, along with variable populations in control and intervention groups leading to high heterogeneity level. The significant reduction of relative risk is supported by two potential processes. First, capsaicin promotes the activation of the TRPV1  receptor which through a cascade effect leads to thermogenesis, fat metabolism and other energy dissipation processes . This wa5To our knowledge, this is the first systematic review and meta-analysis that attempt to identify association between CP consumption and mortality. Statistically significant results of our analysis put forward a rationale indicating an association between lower risk of all-cause, cardiovascular and cancer related deaths and CP consumption .None to declare.


To explore the relationship between ERCC1 rs11615 polymorphism and chemosensitivity to platinum drugs in ovarian cancer by the method of meta-analysis.Pubmed, Web of Science, EMBASE, Cochrane Library, China National Knowledge Infrastructure (CNKI), and China Wanfang databases were comprehensively searched up to September 2020, to identify the relationship between ERCC1 rs11615 polymorphism and chemosensitivity of ovarian cancer. The data was analyzed by Stata 15.0 statistic software.P > 0.05). There were no significant differences in recessive, dominant, homozygous, and heterozygous models. In accordance with a subgroup analysis of Ethnicity, all genotypes were statistically significant in the Asian population. In the allelic, dominant, recessive, homozygous and heterozygous models, the OR was 0.70 (95%CI:0.51 ~ 0.95), 0.20 (95%CI:0.07 ~ 0.56), 0.79 (95%CI:0.63 ~ 1.00), 0.21 (95%CI:0.07 ~ 0.59), 0.19 (95%CI:0.07 ~ 0.54), respectively, while in the Caucasian population, no statistically significant genotype was found.A total of 10 published papers were included, including 1866 patients with ovarian cancer. The results showed that compared allele C at ERCC1 rs11615 locus with allele T, the pooled OR was 0.92 (95%CI:0.68 ~ 1.24, The ERCC1 rs11615 polymorphism is associated with chemosensitivity in patients with ovarian cancer, especially in the Asian population, but not in the Caucasian population. The ovarian malignant tumor is one of the common malignant tumors of female reproductive organs, which is a severe threat to women’s health. Its incidence is second only to cervical cancer and uterine cancer, and its mortality ranks first among gynecological malignant tumors . AccordiNucleotide excision repair (NER) is one of the classic ways to participate in multi-drug resistance in ovarian cancer . As an ePubmed, Web of Science, EMBASE, Cochrane Library, China National Knowledge Infrastructure (CNKI), and China Wanfang databases were searched for literature about ERCC1 rs11615 polymorphism and chemosensitivity of ovarian cancer, from inception to September 2020. The search strategy was as follows: (“ERCC1” OR “excision repair cross-complementation group 1”) AND (“ovarian carcinoma” OR “ovarian cancer” OR “ovarian tumor”) AND  patients with ovarian cancer confirmed by pathological examination; (2) patients received platinum-based combined chemotherapy after operation; (3) genotypes were detected by genomic DNA extracted from tumor tissue or peripheral blood; (4) studies with a large number of cases were selected if the same samples were used in multiple articles; (5) the target gene polymorphism was ERCC1 rs11615 locus or 19007C/T.(1) reviews, letters and conference summaries, etc.; (2) repeated reports, insufficient data, too little reported information, and unavailable literature; (3) studies with Newcastle–Ottawa scale (NOS) scores less than 6.The quality of the included studies was evaluated according to NOS , which wAll literature data were extracted independently by two researchers and finally cross-checked. The following information was extracted from each included literature: author, year of publication, country of the first author, patient ethnicity, total number of cases, chemotherapy regimen, genotype detection methods, and the number of patients with genotype sensitivity and drug resistance.2 ≥ 50%, or P ≤ 0.05, the heterogeneity was considered to exist, and thus a random-effects model (REM) was adopted. If I2 < 50% and P > 0.05, the heterogeneity was considered to be not significant, so a fixed-effects model (FEM) was used for data merging. If there was heterogeneity, subgroup analysis was carried out according to Ethnicity and Hardy–Weinberg Equilibrium (HWE) to explore the source of heterogeneity. Z-test was used to test the significance of the pooled OR value. Funnel plot and Egger’s Test were used to evaluate publication bias. The funnel plot was made by filling and trimming methods. Finally, a sensitivity analysis was conducted to verify the robustness of the research results.The data were analyzed by Stata 15.0 statistical software. The odds ratio (OR) and its 95%CI were calculated as effect size. Q-test was used to assess the heterogeneity across the studies included. If IAccording to the strict inclusion criteria, a total of 10 articles were included , 24–31, 2 = 66.4% (P < 0.05), which indicated significant differences in heterogeneities between studies, so the REM was adopted. The forest plot was shown in Fig. P > 0.05). Ethnic subgroup analysis showed that for the Asian population, OR = 0.70 (95%CI:0.51 ~ 0.95), the difference was statistically significant, and the heterogeneity also decreased. While in the Caucasian population, the difference was not statistically significant. Subgroup analysis of HWE showed that there was no notable decrease in heterogeneity. The symmetry of the funnel plot was general , indicating the existence of the heterogeneity. Hence, a FEM model was adopted. The forest plot was as shown in Fig. P > 0.05). The ethnic subgroup analysis showed that for the Asian population, OR = 0.20 (95%CI:0.07 ~ 0.56), the difference was statistically significant. In the Caucasian population, the difference was not statistically significant. The heterogeneity decreased remarkably. The subgroup analysis of HWE showed that the decrease of heterogeneity was not obvious. The funnel plot was basically symmetrical ), genotype CC + CT was used as the exposure factor, genotype TT as the non-exposure factor, and chemosensitivity as the control. Heterogeneity test showed that Ical Fig. B, and Eg2 = 28.7% (P > 0.05), indicating that the heterogeneity across studies was not significant. Therefore, a FEM was used. The forest plot was as shown in Fig. P > 0.05), which was consistent with the result of the subgroup analysis of HWN. The ethnic subgroup analysis showed that in the Asian population, OR = 0.21 (95%CI: 0.07 ~ 0.59), the difference was statistically noteworthy. In the Caucasian population, the difference was not statistically meaningful. The subgroup analysis of HWE showed that there was no notable decrease in heterogeneity. The funnel plot was basically symmetrical , genotype CC was used as the exposure factor, genotype CT + TT as the non-exposure factor, and chemosensitivity as the control. The heterogeneity test showed that Ical Fig. C, while 2 = 75.1% (P < 0.05), indicating that there were noteworthy differences in heterogeneity across studies. Hence, a REM was adopted. The forest plot was as shown in Fig. P > 0.05). The ethnic subgroup analysis showed that for the Asian population, OR = 0.21 (95%CI: 0.07 ~ 0.59), the difference was statistically significant. In the Caucasian population, the difference was not statistically meaningful. The heterogeneity decreased apparently. Subgroup analysis of HWE showed that the decline in heterogeneity was not significant. The funnel plot was basically symmetrical , genotype CC was used as the exposure factor, genotype TT as the non-exposure factor, and chemosensitivity as the control. The results of the heterogeneity test showed that Ical Fig. D, and Eg2 = 74.4% (P < 0.05), indicating that there was a statistically significant difference in heterogeneity between studies. Therefore a REM was adopted. The forest plot was as shown in Fig. P > 0.05). The ethnic subgroup analysis showed that for the Asian population, OR = 0.19 (95%CI:0.07 ~ 0.54), the difference was statistically meaningful. In the Caucasian population, the difference was not statistically noteworthy. The heterogeneity decreased substantially. While the subgroup analysis of HWE showed that the heterogeneity did not decrease. The funnel plot was basically symmetrical , genotype CT was set as the exposure factor and genotype TT as the non-exposure factor. The results of the heterogeneity test showed that Ical Fig. E, and EgIn consideration of the correlation between ERCC1 rs11615 polymorphism and platinum chemosensitivity of ovarian cancer in the Asian population, a targeted sensitivity analysis in the Asian population was conducted Fig. A-E. The At present, the conclusions upon the rs11615 polymorphism of the ERCC1 gene and the chemosensitivity of platinum drugs in ovarian cancer are not consistent. In order to explore the relationship between them, this meta-analysis was conducted, of which the results showed that there was a correlation between the rs11615 polymorphism of ERCC1 and the chemosensitivity of platinum drugs in ovarian cancer, mainly in the Asian population, but not in the Caucasian population. There are numerous factors that affect the sensitivity of chemotherapy, among which the polymorphism of ovarian oncogene and the chemosensitivity and clinical prognosis of platinum drugs have always been the focus of research , 33. TheP-value of Egger’s Test in both allele model and recessive gene inheritance model was less than 0.05, indicating that there was a certain publication bias in the two genetic models. As to other genetic models, the P values of Egger’s Test were all greater than 0.05, which suggested that there was no significant publication bias in these genetic models. Consequently, it can be concluded that the rs11615 polymorphism of the ERCC1 gene is associated with ovarian cancer patients in the Asian population. Allele C, genotype CC, and CT will increase the sensitivity of ovarian cancer patients in the Asian population to platinum chemotherapy, but some genetic models still need to be explained carefully. Tang et al. [This meta-analysis eventually included ten studies with a total of 1866 patients with ovarian cancer. The results showed that the rs11615 polymorphism of the ERCC1 gene was strongly associated with ovarian cancer patients in the Asian population, and the differences were statistically important in allele, dominant, recessive, homozygous, and heterozygous genetic models. However, no such correlation was discovered in the Caucasian population. In accordance with the ethnicity subgroup analysis, whether in the Asian population or the Caucasian population, the heterogeneity declined significantly. A subgroup analysis was also conducted on whether the genotype frequency of the control group meets HWE, the results of which showed that according to the HWE subgroup analysis, the heterogeneity in each gene model did not notably decrease. The sensitivity analysis results in the Asian population showed that the conclusion in the heterozygous model was robust, but after the allelic, dominant, recessive, homozygous, and heterozygous genetic models were removed, the difference became not statistically significant. In each gene model, the funnel plot was approximately symmetrical. The g et al.  reportedg et al.  reportedg et al.  showed tAt present, our study found that there is a correlation between ERCC1 rs11615 polymorphism and chemotherapy sensitivity of ovarian cancer in the Asian population, but not in the Caucasian population. It may be that there are differences between chemotherapy sensitivity and ovarian cancer patients of various ethnicity. It has been reported that several differences exist in chemotherapy sensitivity to gynecological tumors in different races . In addiInevitably, this study is limited by the following factors. First, the number of literature included was relatively small. After ethnicity subgroup analysis, there were only five studies in the Asian population and the Caucasian population, respectively, which had a certain impact on the robustness of the conclusion. Second, the sample size was relatively small, which was likely to affect the testing efficiency of statistics. Third, there was a certain publication bias in allele model and recessive gene model. Last but not least, in the sensitivity analysis of the Asian population, only the conclusion of the heterozygous genetic model was stable, while the conclusion of other genetic models was unstable to a certain degree.In conclusion, the ERCC1 rs11615 polymorphism is associated with the sensitivity of platinum-based combination chemotherapy in patients with ovarian cancer, especially in the Asian population. Allele C, genotype CC, and CT will increase the sensitivity of ovarian cancer patients to platinum chemotherapy in the Asian population. Nevertheless, no such correlation has been found in the Caucasian population. Given that there are certain limitations in this study, such as small sample size and publication bias, the conclusions of this study still need to be further verified by larger sample size and high-quality clinical research.