Myometrial invasion has been demonstrated to correlate to clinicopathological characteristics and prognosis in endometrial cancer. However, not all the studies have the consistent results and no meta-analysis has investigated the association of myometrial invasion with lymphovascular space invasion (LVSI), lymph node metastasis (LNM), recurrence, and overall survival (OS). Therefore, a meta-analysis was performed to evaluate the relationship between myometrial invasion and clinicopathological characteristics or overall survival in endometrial cancer.A search of Pubmed, Embase, and Web of Science was carried out to collect relevant studies from their inception until June 30, 2021. The quality of each included study was evaluated using Newcastle–Ottawa scale (NOS) scale. Review Manager version 5.4 was employed to conduct the meta-analysis.p < 0.00001), lymph node metastasis (LNM) , and recurrence . Deep myometrial invasion was also significantly related with poor OS via meta-synthesis of HRs in both univariate survival  and multivariate survival . Funnel plot suggested that there was no significant publication bias in this study.A total of 79 articles with 68,870 endometrial cancer patients were eligible including 9 articles for LVSI, 29 articles for LNM, 8 for recurrence, and 37 for OS in this meta-analysis. Myometrial invasion was associated with LVSI (RR 3.07; 95% CI 2.17–4.35; Deep myometrial invasion correlated to positive LVSI, positive LNM, cancer recurrence, and poor OS for endometrial cancer patients, indicating that myometrial invasion was a useful evaluation criterion to associate with clinical outcomes and prognosis of endometrial cancer since depth of myometrial invasion can be assessed before surgery. The large scale and comprehensive meta-analysis suggested that we should pay more attention to myometrial invasion in clinical practice, and its underlying mechanism also deserves further investigation. Endometrial cancer is the most prevalent gynecological malignancy in developed countries  and the Myometrial invasion is defined as the invasion of endometrial cancer cells into myometrium. The depth of invasion is critical to the evaluation of surgical-pathological staging. According to the International Federation of Gynecology and Obstetrics (FIGO) staging system, stage IA includes those tumors with myometrial invasion of less than 50% or without myometrial invasion, and stage IB refers to more than 50% of invasion into myometrium. Although the underlying mechanism of myometrial invasion is still unclear, it is one of critical considerations for the surgery types and therapeutic methods. This is because accumulated evidences show that myometrial invasion is related to LVSI, LNM, recurrence, and OS of endometrial cancer in different reports. However, there are some different sounds, and not all the studies share the similar results. Specifically, there were more patients with positive LVSI in superficial myometrial invasion group when compared to deep myometrial invasion group . In addiCurrently, it is still a mystery whether the above non-uniform results change our previous conclusions and consensus. Until now, there is no aggregated estimate about the relationship between myometrial invasion and LVSI, LNM, recurrence, and OS. This meta-analysis is aiming to further elucidate whether myometrial invasion correlates to LVSI, LNM, recurrence, and OS based on the data available so far. The meta-analysis of multiple clinical studies will provide comprehensive descriptions about myometrial invasion not only from the past to the present but also from clinicopathological characteristics to prognostic value. We hope that the study also could provide us more certainty and confidence in mention and further investigation of myometrial invasion of endometrial cancer.Literature was searched from Pubmed, Embase, and Web of Science from their inception until June 2021. The study published only in English was further considered. The main search terms were formulated as follows: “endometrial cancer”, “endometrial carcinoma”, “endometrial tumor”, “uterine carcinoma”, “uterine cancer”, “endometrial neoplasms”, “myometrial invasion”, “myometrial infiltration”, “clinicopathological factors”, “lymphovascular space invasion”, “lymph node metastasis”, “prognostic marker”, “prognosis”, “overall survival”, “recurrence”, and “relapse”.The study had to meet the following inclusion criteria: (1) the patients only had endometrial cancer; (2) enough data about clinicopathological factors  and/or related information to extract hazard ratio (HR) and standard error (SE) of lnHR for OS; (3) article was published in English. The exclusion criteria included the following terms: (1) reviews, meta-analysis, animal experiments, and case reports; (2) republished articles; (3) incomplete and unpublished studies; (4) the study did not meet the design. Two reviewers independently reviewed the literatures according to the predefined strategy and criteria. The articles were screened with two researchers independently (JW and PX). The disagreements were further settled through discussion and resolved by a third investigator when necessary.Two investigators (JW and PX) were assigned to assess the eligibility of all studies. Moreover, a third investigator (XZ) resolved the disagreements when necessary. The following information from each study was extracted: first author, publication year, the region of the study population, the number of participants, design type. For LVSI, LNM and recurrence, and the numbers in case and control groups were extracted respectively. For OS, HR estimate with 95% confidence interval (CI) for OS was extracted. The quality of included studies was assessed using Newcastle–Ottawa scale (NOS) scale, and the score of the quality ranged from 0 to 9.p ≤ 0.1, I2 > 50%) was shown. Publication bias was evaluated by the shape of funnel plot. Statistically significant difference was pointed out when a p value was less than 0.05.Version 5.4 software of Review Manager was applied for this meta-analysis. Risk ratio (RR) with 95% CI was pooled to investigate the association between myometrial invasion and clinicopathological features . HR with 95% CI was combined to study the effect of myometrial invasion on OS. The HR was extracted directly when the HR with 95% CI was reported. SE was calculated using the equation: SElnHR = (lnUpperCI − lnLowerCI)/3.92 . A randoThe predefined search strategy identified 1,385 records. After screening of titles and abstracts, 1,058 records were excluded including 35 non-English papers, 18 duplicated records, 208 review/meta/letter/abstract, 19 animal studies, and 778 irrelated literature. Full text of 327 articles was assessed, and 248 records were excluded including 8 studies with the same included patients, 11 basic research, and 229 articles without adequate data. Finally, 79 studies of total 68,870 patients were eligible . The risk ratio, which was expressed as >1/2 group versus <1/2 group, was 3.07  . The risk ratio was 4.45  , and fixed-effects model was used. Myometrial invasion was significantly associated with the recurrence of endometrial cancer since the risk ratio was 2.06   showed that the group with deep myometrial invasion was related with a higher risk of OS than the group with less than 1/2 myometrial invasion. Therefore, deep myometrial invasion is associated with poor survival in endometrial cancer.Thirty-seven studies including 9,416 patients examined the association between myometrial invasion and OS in endometrial cancer. The pooled HRs of all-cause mortality with >1/2 myometrial invasion compared to <1/2 myometrial invasion were evaluated using random-effects model, and the results are presented in Funnel plot was applied for the assessment of publication bias in the literature, and tests for funnel plot asymmetry were applied only when there were at least 10 studies included in a meta-analysis. The shape of the funnel plot for the included 29 studies on the association between myometrial invasion and LNM was not significantly asymmetrical, indicating that there was no significant publication bias  or OS and conducted this meta-analysis.The presence of LVSI is significantly associated with pelvic and paraaortic lymph node metastasis, recurrence, and poor prognosis , 87. As Seventy-nine studies with a total of 68,870 endometrial cancer patients were finally included for this meta-analysis. Among them, nine studies with a total of 28,904 endometrial cancer patients were for LVSI. The pooled result showed patients with deeper myometrial invasion of endometrial cancer into myometrium (>1/2) were more prone to LVSI. As for LNM, 29 studies including 31,262 endometrial cancer patients were eligible for analysis, and the results demonstrated that deeper myometrial invasion is associated with the tendency of LNM in endometrial cancer. Furthermore, myometrial invasion was significantly associated with the recurrence of endometrial cancer according to the meta-analysis of eight studies including 1,649 patients. Since LVSI, LNM, and recurrence are independent prognostic factors for endometrial cancer patients, myometrial invasion would also be a prognostic factor. As it turned out, the group with deep myometrial invasion was related with a greater risk of OS than the group with less than 1/2 myometrial invasion based on not only univariate survival analysis but also multivariate survival analysis. Therefore, the results indicate that myometrial invasion is associated with LVSI, LNM, recurrence, and OS with much more confidence. Combined with preoperative assessment of the depth of myometrial invasion, now we know more information in regard to LVSI, LNM, recurrence, and OS of these patients before surgery, which suggests that we should especially pay more attention to myometrial invasion in clinical practice, and its underlying mechanism also deserves further investigation.Potential limitations exist in this study, and meta-analysis without the classification of endometrial cancer is the obvious one. In the past, dualistic classification is the leading theory for the classification, which divides endometrial cancer into type I and type II tumors . AccordiApart from classification, other potential limitations still exist in this study: (1) The data from the included studies were from the published articles instead of the original information of individual patient; (2) most included articles are the retrospective studies, and the evidence level is lower than that of prospective randomized clinical trial; (3) one of inclusion criteria is that article was published in English and negative results not being reported, which increase the risk of publication bias; (4) the number of included studies is relatively small, especially for LVSI and recurrence, which may cause biased results; (5) the heterogeneity of aggregated results was significant, and the random-effects model was applied.In summary, a large scale and comprehensive meta-analysis of the association between myometrial invasion and other clinicopathological characteristics and prognosis is provided in the present study. Our results show that myometrial invasion is associated with LVSI, LNM, recurrence, and OS, indicating that deep myometrial invasion is a useful evaluation criterion to associate with poor clinical outcomes and prognosis in endometrial cancer patients.The original contributions presented in the study are included in the article/supplementary material. Further inquiries can be directed to the corresponding author.JW, PX, and XZ: conceptualization. JW, PX, and XY: data curation and original draft writing. QY, XX, and GZ: statistical analysis. JW and XZ: manuscript review and editing. All authors contributed to the article and approved the submitted version.This study was supported by National Natural Science Foundation of China (Grant numbers: 81802591 and 81974225) and National Key R&D Program of China (Grant number: 2017YFC1001202).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.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


The aim is to comprehensively and accurately assess potential relationships between single nucleotide polymorphisms (SNP) and lung cancer (LC) risk by summarizing the evidence in systematic reviews and meta-analyses. This umbrella review was registered with the PROSPERO international prospective register of systematic reviews under registration number CRD42020204685. The PubMed, Web of Science, and Embase databases were searched to identify eligible systematic reviews and meta-analyses from inception to August 14, 2020. The evaluation of cumulative evidence was conducted for associations with nominally statistical significance based on the Venice criteria and false positive report probability (FPRP). This umbrella review finally included 120 articles of a total of 190 SNP. The median number of studies and sample size included in the meta-analyses were five  and 4 389 , respectively. A total of 85 SNP (in 218 genetic models) were nominally statistically associated with LC risk. Based on the Venice criteria and FPRP, 13 SNP (in 22 genetic models), 47 SNP (in 99 genetic models), and 55 SNP (in 94 genetic models) had strong, moderate, and weak cumulative evidence of associations with LC risk, respectively. In conclusion, this umbrella review indicated that only 13 SNP (of 11 genes and one miRNA) were strongly correlated to LC risk. These findings can serve as a general and helpful reference for further genetic studies. Lung cancer (LC) is associated with high morbidity and mortality rates and, thus, remains a serious threat to human health . LC is gMolecular epidemiological and experimental studies have shown that genetic variations play important roles in the occurrence of LC . A singlDong et al. evaluated the results of meta-analyses and pooled analyses along with the false positive report probability (FPRP) to summarize the genetic susceptibility to cancer and found only 11 significant associations between genetic variations and LC risk . MarshalWe conducted an umbrella review, which systematically collected and evaluated systematic reviews and meta-analyses of a specific research topic . The umbEligible systematic reviews and meta-analyses published until August 14, 2020 were retrieved from the PubMed, Web of Science, and Embase databases with a combination of subject headings and free terms, as detailed in The inclusion criteria for article eligibility were: 1) systematic reviews or meta-analyses with quantitative synthesis; 2) investigations of the association between SNP and LC risk; 3) inclusion limited to observational studies, while excluding cross-sectional studies; 4) case-control studies or genome-wide association studies (GWAS) included in the meta-analyses that provided the number of cases and controls, and cohort studies included in the meta-analyses that provided the number of cases and population participants; 5) providing the genotyping data or specific relative risk estimates  with the 95% confidence interval (CI) of each included study; 6) included at least three studies; and 7) the article was written in English.The exclusion criteria of eligible articles were: 1) included studies whose subjects were non-human, or studies without cancer-free controls; 2) included family-based studies; 3) investigations of variants with ranges greater than one SNP; 4) evaluation of the diagnosis, survival, or recurrence of LC; 5) meta-analyses or systematic reviews based on individual data; and 6) unpublished articles, published articles in abstracts only, letters to editors, and editorial comments.If there was more than one eligible meta-analysis of the same SNP, the most recently published one  with the corresponding data described in inclusion criteria 4) and 5) were retained because the most recent meta-analysis usually had the largest sample size  . If an ap) value of the Hardy-Weinberg equilibrium (HWE) test for each of the included studies.Two investigators separately extracted data from the eligible systematic reviews and meta-analyses and a dedicated investigator conducted quality control and resolved inconsistencies. For each eligible article, the extracted data included 1) the name of the first author, 2) year of publication, 3) examined SNP, 4) gene name, 5) the number of included studies, 6) genotyping data or specific relative risk estimates  with the 95% CI for each of the included studies (genotyping data was preferred), 7) epidemiological design  of each study, 8) the number of cases and controls (for case-control studies and GWAS) or the number of cases and population participants (for cohort studies) of each study, and 9) the probability , the homozygote comparison model (model 2), the dominant model (model 3), the recessive model (model 4), and the allele model (model 5) .p < 0.05 (random-effects model) was considered nominally statistically significant for each meta-analysis   the larger studies had a more conservative effect size than the random-effects meta-analysis .For SNP with nominal statistical significance, four methods were used to assess bias. First, for nominally statistically significant relationships, we examined whether the relationships were lost by excluding the first published studies . Second, < 0.05) . Third, analysis . Fourth,analysis . Brieflyhttps://www.ncbi.nlm.nih.gov/snp/). MAF usually refers to the frequency of alleles that are uncommon in a given population. Finally, an association with a rating of AAA was considered strong, while a rating of C for any of the three items was considered weak. All other ratings were considered moderate.The cumulative evidence of SNP with nominal statistical significance was further evaluated. First, the strength of the evidence, as an indicator of epidemiological credibility, was evaluated using the Venice criteria  that havThe FPRP is a Bayesian prophylactic against false reports of significant associations. The FPRP was calculated with the Excel spreadsheet on the Wacholder website . For FPROf the 3,065 records initially retrieved from the PubMed, Web of Science, and Embase databases, 1,774 (57.88%) were retained after removing duplicates. Finally, 120 articles were included in the umbrella review , which rn = 89, 74.17%) were published since 2015 . With respect to the epidemiological design, most of the meta-analyses (91.58%) were synthesized by case-control studies as well as nested case-control studies. Some meta-analyses (8.42%) also included the training set and validated set of GWAS. Included subjects were from many regions of the world, as shown in The basic characteristics of the included articles are summarized in The quality of all included articles was assessed and the results were listed in A total of 85 SNP were nominally statistically associated with LC risk in at least one genetic model. Of these 85 SNP, 83 were located on 54 genes and two were located on two miRNAs. The median number of studies included in the meta-analyses was six . The median number of cases, controls, and total sample size included in the meta-analyses were 3,173 , 3,578 , and 7,016 , respectively .CLPTM1L gene was rated as strong evidence. A total of 32 genetic models of 19 SNP and 182 genetic models of 79 SNP were rated as moderate and weak evidence, respectively.The Venice criteria were used to assess the strength of evidence . Only rsThe Venice criteria and FPRP were combined to more accurately evaluate the cumulative evidence . There wA total of 148 SNP were not nominally statistically significant in at least one genetic model . Of thesSensitivity analyses of the corresponding models of SNP with the strong cumulative evidence  failed thttp://bioinfo.life.hust.edu.cn/web/GSCALite/), eight (72.7%) of these 11 genes  might be involved in one or more pathways related to LC  had moderate cumulative evidence, and 55 SNP (in 94 genetic models) had weak cumulative evidence. In general, the results of this umbrella review were not in very good agreement with the review by ed to LC .AXIN2 (axis inhibition protein 2) gene, also known as AXIL and ODCRCS, is a negative regulator of the Wnt/β-catenin signaling pathway that may play an important role in tumorigenesis (AXIN2 coding region (exon1) at 17q24.1, suggesting that rs2240308 may significantly influence AXIN2 gene expression. In the present umbrella review, rs2240308 was associated with susceptibility to LC, as suggested by the strong cumulative evidence in the dominant model. As compared with the CC genotype, the TT + CT genotype was associated with a significantly reduced risk of LC. However, the sample size of rs2240308 in the meta-analysis was relatively small, thus further investigations are necessary.The igenesis . SNP rs2CHRNA3 (cholinergic receptor nicotinic α3) gene, also known as LNCR2, PAOD2, BAIPRCK, and NACHRA3, encodes the α3 nAChR (nicotinic acetylcholine receptor) subunit. A study by Paliwal et al. demonstrated that depletion and restoration of CHRNA3 expression induces and resists cell apoptosis, respectively  gene, also known as CRR9, is a LC risk candidate gene that was found to be overexpressed in human lung tumor cell lines and lung tumors  gene, also known as AHH, AHRR, CP11, CYP1, CYPIA1, P1-450, P450-C, and P450DX, encodes a phase I enzyme that adjusts the metabolic activation of important tobacco procarcinogens, such as polycyclic aromatic hydrocarbons and aromatic amines, and might influence susceptibility to LC by regulating the metabolism of environmental carcinogens  gene encodes a DNA repair protein that is vital to the repair of DNA damage induced by alkylating agents. Studies have demonstrated that MGMT plays an important role in the pathogenesis of cancers and might be a good biomarker candidate for early cancer detection  gene, also known as NBN, ATV, NBS, P95, AT-V1, and AT-V2, has an important influence on the cellular response to DNA damage and maintaining chromosomal integrity, which might influence oncogenesis  gene, also known as TP2, TRT, CMM9, EST2, TCS1, hTRT, DKCA2, DKCB4, hEST2, and PFBMFT1, encodes the TERT protein, which is the catalytic subunit of telomerase and plays a vital role in the maintenance of telomere stability  gene, also known as XP1 and XPAC, encodes the XPA protein, which is a DNA damage recognition and repair factor. As a zinc finger DNA binding protein, XPA is essential to nucleotide excision repair. So, a mutation to the XPA gene might be involved in oncogenesis  gene, also known as AT1, ATA, ATC, ATD, ATE, ATDC, TEL1, and TELO1, is a cancer-susceptibility gene that encodes the ATM protein, which takes part in the identification and repair of DNA damage and cell cycle regulation. Thus, a mutation to the ATM gene might induce not only multiple system dysfunction, but also a concomitant increase in susceptibility to LC  gene, also known as CPE1, CYP2E, P450-J, and P450C2E, encodes the CYP2E1 protein, which is an ethanol-inducible enzyme. CYP2E1 can metabolically activate various carcinogens, including benzene and N-nitrosamines in tobacco, and thus might play a vital role in the development of LC  gene, also known as RCC and SCAR26, encodes a DNA repair protein that can interact with DNA components at damage sites to fix DNA base damage and single-strand breaks  was strongly associated with an increased risk of LC. A previous study reported that miR-1262 on 1p31.3 may suppress the proliferation of LC cells . SNP rs1APEX1 (rs1130409), COX-2 (rs5275), EPHX1 , ERCC1 (rs11615), ERCC5 (rs17655), FASL (rs763110), MTHFR (rs1801131), NQO1 (rs1800566), TP53 , XPC (rs2228001), XRCC1 (rs25489), and XRCC3 (rs861539). According to the calculation results obtained with Quanto 1.2.4 software (https://preventivemedicine.usc.edu/download-quanto/), 10,000 subjects provided approximately 80% statistical power if the incidence of LC was 200 per 100,000, OR was 1.15, the genetic model was the dominant model, and the MAF was 0.1. Therefore, further investigations of these 14 SNP might not be very productive.In addition, 81 of the SNP identified in this umbrella review were not significantly correlated to LC risk in any of the five genetic models. Of these 81 SNP, 14 SNP on 12 genes had a sample size of more than 10,000, which included There were certain advantages to this umbrella review. This is the first umbrella review that extracted data, rather than the results, of systematic reviews and meta-analyses to calculate and evaluate the associations between SNP and LC risk. Moreover, the combined use of the Venice criteria and FPRP increased the reliability of the assessment results. However, there were some limitations. First, subgroup analysis stratified by ethnicity was not conducted because if an umbrella review of a specific ethnicity was included, the most recent meta-analysis referring to this ethnicity must be screened again. Hence, another umbrella review of a specific ethnicity is planned in the future. Second, the quality of included meta-analyses varied to a certain extent, which might lead to data credibility issues. Third, grey literature was not included in this umbrella review.In conclusion, this umbrella review found strong cumulative evidence of associations of 13 SNP (of 11 genes and 1 miRNA) with LC risk, which provides important references for future studies on the relationships between SNP and LC risk.


Cancer and heart failure (HF) are two leading global causes of morbidity and mortality and bring a huge disease burden to the world. Although cancer and HF commonly coexist, HF is often considered as a complication secondary to the cardio-toxicity of anti-cancer treatments and much attention has been paid to HF incidence among cancer patients. On the contrary, cancer development as a consequence of HF, has been far less studied. However, in recent years, accumulating evidence has suggested HF and cancer share lots of common risk factors and have various similarities. Inflammation, obesity, oxidative stress, diabetes, hypertension, and others are all contributors to the development of HF and cancer , 2. It s. Although a previous meta-analysis of only 4 studies demonstrated a correlation between HF and increased cancer risk  and random-effects model was deployed for high heterogeneity among studies (I2 ≥ 50%).Two investigators independently reviewed published studies in PubMed, EMBASE and Web of Science databases from their inception to October 2021 using the search strategy that included the terms for “HF” and “incident cancer.” Then, we manually searched for additional studies using references of selected retrieved articles to identify other possible studies. There was no limitation on language. Studies were included according to the following inclusion criteria: 1. cohort study; 2. HF was diagnosed before cancer incidence; 3. hazard ratio (HR) or equivalents with 95% confidential interval (CI) was reported. We excluded experimental studies, clinical trials, cross-sectional studies, case–control studies, reviews, commentaries, letters and conference abstracts. Data extraction was completed independently by two investigators, using a predefined data extract form. Disagreements were resolved by discussion among all investigators. Information on study characteristics, participants' features, exposure and outcomes were extracted. If both the crude and adjusted values were provided, we only extracted the adjusted values. The analysis was performed using Stata12.0 software . Fixed-effects model was used for low heterogeneity among studies (n = 126) or article types (n = 296) or because the title and abstract did not meet the inclusion criteria (n = 288). The retrospective study by Sakamoto et al.  (P = 0.068) (P = 0.139) or breast cancer   without statistical significance (= 0.056) . Conside= 0.068) . As to t= 0.166) .Our updated meta-analysis summarized all cohort studies presently available and demonstrated that there was no significantly increased risk of incident cancer among HF patients. In view of the high heterogeneity among published studies, bias may have existed. As presented in Kwak's study , an abruIn conclusion, the available evidence still could not clearly identify HF as a new risk factor for incident cancer. Results should be interpreted with caution due to the high heterogeneity among studies. Further research is needed to elucidate the conflicting clinical data and uncover the underlying mechanisms.XZ contributed to conceptualization, data collection, and writing. NL contributed to data collection, analysis, and editing. YZ contributed to analysis. JZ contributed to the conceptualization, illustration, and editing of the manuscript. All authors approved the submitted version.This work was supported by National Natural Science Foundation of China (82104588) and Pyramid Talent Program of Changzheng Hospital (YQ662) to JZ.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.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


The objective of the present article was to qualitatively and quantitatively review the association between chronic mechanical irritation and oral squamous cell carcinoma (OSCC). PubMed, SCOPUS, and Web of Science databases were searched using the keyword combinations “chronic trauma and oral squamous cell carcinoma; chronic irritation and oral squamous cell carcinoma; chronic irritation and oral cancer; and chronic trauma and oral cancer.” Duplicates and irrelevant articles were excluded after the title and abstract screening. The full texts of the remaining articles were assessed using selection criteria. A total of 375  articles were screened, and 343 duplicates and irrelevant articles were excluded from the study. Only 9 of the remaining 32 articles met the selection criteria and were included in the qualitative analysis. Buccal mucosa and tongue, being highly prone to chronic irritation through the dental prosthesis, were the common sites for OSCC. Edentulous subjects with ill-fitting dentures were at a high risk of developing chronic irritation associated-OSCC. According to the Joanna Briggs Institute of risk assessment, eight of the nine included studies had a low risk of bias. The quantitative analysis showed a significant association (p < 0.00001) between the chronic oral mucosal irritation and OSCC with an overall risk ratio of 2.56 at a confidence interval of 1.96-3.35. Chronic oral mucosa irritation has a significant association with OSCC, and the nature of association could be that of a potential co-factor (dependent risk factor) rather than an independent risk factor. Oral squamous cell carcinoma (OSCC) the sixth most common neoplasm of the head and neck regions accounts for 80-90% of malignancies of the oral cavity . Around The multifactorial etiology of OSCC includes tobacco, alcohol, and betel quid with and without added tobacco as the major risk factors . In addiOne such poorly-explored factor is chronic mechanical irritation (CMI). There have been cases reported with OSCC developing in an oral site with a history of CMI. The CMI may result from poor oral hygiene,  poor denBefore exploring into adapting prophylactic measures, it is vital to confirm the association between CMI and OSCC. The lack of conclusive evidence for the same can be largely attributed to the presence of confounding factors in most of the studies. The inclusion of inflammation as the seventh hallmark of cancer since 2009  had led The International prospective register of systematic reviews (PROSPERO) was searched for systematic reviews assessing the potential associating between OSCC and CMI, following which the present review was submitted for registration. Preferred Reporting Items for Systematic Reviews and Meta-Analysis Protocols (PRISMA) were strictly adhered to Shamseer et al. .The review was conducted in 3 steps:· Keyword combinations “Chronic Irritation and Oral Squamous Cell Carcinoma, Chronic Irritation and Oral Cancer, Chronic trauma and Oral Squamous Cell Carcinoma and Chronic trauma and Oral Cancer” were used in SCOPUS, PubMed, and Web of Science databases to identify potential articles· The titles and the abstract of the identified articles were screened by AAG and SK for relevance and potential duplicates· The full text of the article selected through screening was assessed by AAG and SK using the following selection criteria:Original research studies published in the English language assessing the potential association between CMI and OSCC.Reviews, short articles , case reports, experimental animal models, and articles published in languages other than English.Step 2 and 3 were conducted by two reviewers . Kappa coefficient (k) was calculated for steps 2 and 3 to determine the inter-observer reliability.Vital data including the aim of the study, the sample characteristics, nature of trauma, sample size, the results including the odds ratio, and the inference were retrieved from the included studies.Joanna Briggs Institute’s methodology for cross-sectional studies and case control studies was used to assess the risk of bias in the included studies ,28.nd and 3rd step of the review process indicating a good overall inter-observer reliability. A total of 375 articles including 126 from PubMed, 152 from Scopus, and 97 from Web of Science were retrieved using the keywords. Screening the titles and abstracts of the identified articles revealed that 343 articles were either duplicate or were not related to the topic of interest and thus were excluded from the study. The full text of the remaining 32 articles was assessed using the selection criteria. Only 9 fulfilled the eligibility criteria and were included in the systematic review. Cross references of these 9 articles were checked manually and another four articles were included which gave a total count of 13 articles that were included in the systematic review. The workflow of systematic review has been summarized in Of the studies included in the review, 3 were from Argentina ,29,30 anDespite using different research models and designs, all the included studies reported a significant association between CMI and OSCC. Most lesions occurred on the buccal mucosa and tongue which are the most common sites to be injured by dental, functional, and prosthetic factors. A higher percentage of edentulism and ill-fitting dentures have been associated with OSCC . In non-p < 0.00001) with OSCC , cervical carcinoma with chronic HPV infection, colorectal carcinoma with inflammatory bowel disease, and carcinoma of the lung with inflammation of lung induced by smoking . ChronicThe narrative review by Piemonte and Lazos used the Bradford Hill Criteria to provide an in-depth assessment of the association between CMI and OSCC. Given the conflicting data, the authors concluded by citing the various shortcomings of their included studies and suggested that future studies provide a clear definition for CMI. The authors had also advised recording the factors that could be considered as triggers for CMI including dental, prosthetic, and functional, especially in OPMD, OSCC cases. The need for exploring the biological pathways of CMI associated with OSCC was also emphasized . The preHence, future research with clear study design, appropriate sample size, and appropriate adjustment of the statistical analysis according to other parameters has to be conducted to establish the role of CMI in cancer progression.The present article qualitatively and quantitatively reviewed the association between CMI and OSCC. Studies with different study designs have shown a common consensus in the associating CMI with OSCC. However, due to the presence of confounding factors and the lack of long-term prospective human studies, it was not possible to assess CMI as an independent risk factor for OSCC. Based on the meta-analysis, CMI has a significant association with OSCC, but the association could potentially be of a co-factor (dependent risk factor) rather than an independent risk factor. Further multicenter prospective clinical studies including patients without known confounding (risk) factors would allow the assessment of CMI as a potential independent risk factor for OSCC.


Introduction: The primary studies demonstrated that fabellar prevalence (FP) varied with ethnic and geographical distribution. Osteoarthritis (OA) and age-related degeneration have a significant association with FP. The prevalence of OA worldwide was doubled with life expectancy. Increased life expectancy has increased exposure to OA and age-related degeneration which could be a possible reason for the rise of FP. The analysis was conducted to provide insight about FP in respect to geographical, ethnic, sex, and laterality distribution.Methodology: Eighty-six studies were included which have data from 34,733 knee joints. Fifty radiological studies were consisting of 27,293 knees and 36 cadaveric studies had the data of 7,440 knees of dissected specimens, respectively. The prevalence, Odds, and rate ratios were calculated for aging, osteoarthritis, and ethnic variation. Results: The worldwide FP was 25% . The prevalence of fabella was found to be higher in cadaveric studies (32%) than radiological studies (19%) with significant heterogeneity. The FP was 16-18% till 1950 which was doubled by 2020 (35%). The FP in OA knee was 51% which was thrice of baseline. The tendon of the lateral head of gastrocnemius has a small fibrocartilaginous or ossified nodule. This nodule is a sesamoid bone named fabella which develops inside the tendon after 8-12 years of age . The ossThe fabellar prevalence (FP) is variable, which depends upon the methodologies, and racial or geographical distributions. Berthaume et al.  claimed The inclusion criteria for this meta-analysis were the published articles, conference abstracts, unpublished studies with obtainable data, and electronic mails used to gather unpublished data from authors of published articles. The case series or literature review were included in the analysis if sample sizes were mentioned. Their references may be utilized to explain the findings. Both cadaveric and radiological studies were included. The data of radiographs, CT, and MRI scans were considered.The mean prevalence mentioned in the textbooks or published literature without sample references was excluded. If the authors of any radiographic study were confused between popliteal artery calcification and fabella, then such data were excluded from the analysis. The data of USG and PET excluded from analysis due to poor sensitivity. The literature associated with mammalian prevalence other than human was excluded based on the titles, abstracts, and full articles. Any reports or case series were excluded if the sample size not mentioned. The systemic search was conducted in the primary database like Medline, Embase, Pubmed, Ebsco, Google Scholar, Ovid database, AUSPORT, and Cochrane library till June 2020. The print or online journals of anatomy, orthopedics, sports or biomechanics, morphology or anthropology science, radiology, and radiotherapy were thoroughly investigated. The search strategy consisted of MeSH terms in different strings of permutations or combinations with Boolean operators, e.g., prevalence, fabella, osteoarthritis, aging, and knee pain.Selection of studies and assessment of the risk of biasThe studies collected from search strategies were shortlisted with the help of Reyyan QCRI app. The data of search strategies were imported to Reyyan QCRI and shortlisted based on inclusion and exclusion criteria and reading the abstract. A total of eighty-six studies were found suitable for further assessment Table . The fulData extraction and management 2 statistics, Cochrane Q, and Kendall Tau. Random effect model was adopted if i2 was more than 50%, in place of the fixed-effect model. Q Cochrane was computed along with the P-value. Sensitivity and cumulative analysis were also performed. The cumulative meta-analysis was performed in order to evaluate the association prevalence of fabella with osteoarthritis and aging. The subgroup analysis was conducted to compute the rate according to ethnicity, mode of study, type of population, sex, and laterality. Regression analysis was carried out to examine the relationship between outcome and confounders . Finally, the odds ratio (OR), rate ratio (RR), and rate difference (RD) were computed. Funnel plots measured the publication bias. For the funnel plot, the logarithmically transformed odds ratio plotted against the standard error of each study. Additionally, the publication bias was measured with the help of Egger's linear regression test, Begg and Mazumdar's rank correlation test. Rosenthal Fail-Safe Number (FSN) was generated to refute the file drawer effect.The information of publishing year, population characteristics, mode of investigation , sample size , and number of fabellae were extracted. The distribution of gender, age, laterality, and ethnicity were also noted. The contingency tables were prepared separately and the event rate with a 95% confidence interval was computed. Age, sex, and laterality were considered confounding factors, so the rate estimations were done separately. The unit of analysis for the prevalence of fabella was the number of fabella in 100 knees. The prevalence of fabella was measured with the help of ProMeta 3. The rate estimate for ossified fabella was measured with Revman 5.3 . The effect size and standard error for each study were computed with ProMeta 3. The heterogeneity statistic was determined as iDescription of studies The included publications were observational - cross-sectional, prospective, and retrospective studies. A total of 122 studies were identified by online research and two studies were found in conference proceedings. Due to duplicate titles or abstracts found during searches, four articles were eliminated. Thirty-six studies were excluded for being non-human studies or case reports and series, reviews-based abstract or title evaluation. A study was included after the consensus of both authors, and the third author was consulted in contradicting opinion. For the meta-analysis, 86 studies that dealt with the prevalence of fabella between 1875 and 2020 based on the abstract and full-text analysis were included in this study  Figure . The exiFP in Different Age-Group and OsteoarthritisThe FP in 0-10 years of the subject was 0.98% and it was increasing with age. The FP was 33.84% in more than 70 years of the aged population  and Asian mongoloid populations . The Oceanian population had the highest FP which would be an over-estimation with a wide confidence interval due to fewer studies Table . The AfrSex and Laterality Distribution Table FP was 27% in males and 24% in females. Bilateral occurrence of fabella was more common than unilateral occurrence. The laterality distribution was estimated per hundred subjects. The bilateral and unilateral occurrences were 61% and 27%, respectively. In the case of unilateral distribution, the right and the left side had equal distribution.Methodological DistributionThe FP in dissected specimens was 32%. The FP in CT scan, MRI, and X-ray studies were 35%, 28%, and 19%, respectively  and the result of Egger's regression test, some publication bias was suspected (P=0.02). But, Begg and Mazumdar's rank correlation test (P=0.126) refuted the asymmetrical funnel plot because of publication bias and it was possibly due to sample variation.Fabellar presence is decided by the genetic constitution as seen in near-human species. The formation of fabella takes place in intrauterine life as a cartilaginous nodule. The cartilaginous nodule is derived from pre-cartilaginous condensations of fibroblast named as a sesamoid precursor in the gastrocnemius tendon during the embryonic period. Hox gene and genes encoding for the TGF-beta family played a crucial role in the sesamoid precursor. Drachman and Sokoloff studied the development of sesamoids in embryonic chicken. The above experiment had explained the fabella development not only determined by the genetic constitution but also by a biomechanical stimulus which led to epigenetic changes .The present study demonstrated that the prevalence of fabella was 25% in 34,733 knees in over 150 years. The FP since the last two decades is estimated at 35% which was 16-18% in 1875-1950. The current meta-analysis has shown a doubling of FP. Berthaume and Bull . If we pFabella is exposed to such changes due to its vicinity to the knee joint. Such changes contribute to alteration in the shape and size of fabella Figure . The enlLimitation of studyDespite the fact that this study offers useful information about the prevalence of fabella, and its relation with OA, the main limitation of this study is the inadequacy of information regarding weight, height, and BMI. The studies used in the assessment were of various ethnicities and methodologies, leading to even greater heterogeneity. Height, gender, physical habitus or profession, and BMI all seem to be potential confounders in the pooled estimates. Prevalence estimates can be influenced by sample variance.This study demonstrated the doubling of FP in the last 70 years. OA knee has three times FP of baseline. The incidence of fabella is higher in the Oceanian and Mongoloid populations, male participants, and on the right side of the knee, which is consistent with previous findings. Bilateral occurrence of fabella is far more common than unilateral occurrence, which corresponds to OA distribution. The valgus positioning of the knee, along with obesity, causes lateral compartment OA in the knee, which can raise the FP. The further scope of the study is to evaluate the prevalence of fabella in severe osteoarthritis with and without valgus knee alignment. The stratification of the proposed study would include a final estimation of FP based on the biomechanical load on the knee, as well as aid in understanding the function of fabella in knee biomechanics.


A synthesis of 40 years of epidemiologic studies of farming and brain cancer that updates two previous meta-analyses finds that farming with its potential for exposure to chemical pesticides is associated with an increased risk of brain cancer.Twenty additional years of epidemiologic literature have become available since the publication of two meta-analyses on farming and brain cancer in 1998. The current systematic literature review and meta-analysis extends previous research and harmonizes findings. A random effects model was used to calculate meta-effect estimates from 52 studies (51 articles or reports), including 11 additional studies since 1998. Forty of the 52 studies reported positive associations between farming and brain cancer with effect estimates ranging from 1.03 to 6.53. The overall meta-risk estimate was 1.13 , suggesting that farming is associated with a 13% increase in risk of brain cancer morbidity or mortality. Farming among white populations was associated with a higher risk of brain cancer than among non-white populations. Livestock farming  was associated with a greater risk compared with crop farming . Farmers with documented exposure to pesticides had greater than a 20% elevated risk of brain cancer. Despite heterogeneity among studies, we conclude that the synthesis of evidence from 40 years of epidemiologic literature supports an association between brain cancer and farming with its potential for exposure to chemical pesticides. While overall cancer death rates in the United States (US) have declined between 1999 and 2018, deaths from primary cancers of the brain and central nervous system (CNS) have increased during 2014 and 2018 . For theThe epidemiology of primary brain cancers, which include meningiomas and gliomas, is largely inconclusive ,6. In thThe agriculture industry, and specifically farming occupations, utilize chemical pesticides and fertilizers, many of which have been assessed for their potential as human carcinogens ,9. ConceAs the world’s population is projected to double by 2050, farmers are expected to continue to use pesticides as important tools among available technologies  to contrUsing PubMed and Agricola databases, a comprehensive literature search was conducted to identify epidemiologic studies evaluating the relationship between farming and brain cancer that were published in English between 1 January 1997 and 1 August 2019 in order to locate studies that had been published after the two previous meta-analyses . The folEpidemiologic studies of adult populations that included farmers .Morbidity or mortality from brain cancer or cancers of the central nervous system was a reported outcome.Measures of effect were estimated in the study, or data were available that allowed for the calculation of a relative risk estimate , Standardized Mortality Ratio (SMR), Proportionate Mortality Ratio (PMR), Relative Risk (RR), Odds Ratio (OR), Proportionate Cancer Mortality Ratio (PCMR), Mortality Odds Ratio (MOR)) and a measure of variability ).Studies were retained if they met the following criteria:When multiple papers were available on the same study population , the most recently published article or most comprehensive analysis on brain cancer in farmers was selected.Two authors (BL and PO) independently reviewed each study to abstract data using a standard format. Information was collected and summarized on each study as follows: publication year (or date of report if not published); source identified  for more than one reference group, such as with a national population and a regional population, we used the latter.When risk estimates were presented for both industry  and occupation , we used the latter and, when available, selected risk estimates associated with “farming” as an occupation. If risk estimates for multiple farming occupations  were presented, we pooled estimates and, when possible, abstracted those which specifically attributed to crop  or livestock  farming. When a study made distinctions between “farmers” and “other agricultural occupations” or “forestry workers and fishermen” in their reporting of risk estimates, we extracted those for “farmers”.When risk estimates were reported for outcomes that included “diseases of the nervous system”, “all nervous system cancers”, and “brain cancer”, we extracted those specific to “brain cancer”. When studies reported separate risk estimates for more than one histological type of brain cancer, or for malignant and benign neoplasms, we pooled estimates across the brain cancer subtypes.When a study reported more than one measure of risk , we prioritized our abstraction as follows: RR, OR, SMR/SIR, PMR/PCMR. If a study reported crude and adjusted risk estimates, we used adjusted measures. If 95% confidence intervals were not reported and approximate CIs could not be calculated with data provided by the study, we used the 90% CIs.When a study reported effect measures for two (or more) separate population cohorts and reported risk estimates for these subcohorts individually, we used these risk estimates as separate observations, i.e., as if they were two (or more) studies. If the study did not provide data to be able to calculate effect measures on the entire cohort, we used the data for the portion of the cohort that was available in our calculations.We handled other alternatives that arose through these rules using a series of sensitivity analyses (see below), which allowed us to examine the influence on the meta-effect measure by including individual effect measures calculated with different reference populations or adjusted for covariates.Two authors (N.M.G. and P.O.) subjectively evaluated the quality of each study based on the completeness of the methods described and results presented. We considered how well farming and exposures were characterized, whether studies provided specific industry or occupation codes, and how historical employment was addressed. In addition, we evaluated how brain cancer was defined , how cases were identified , and whether there was histological confirmation of the cancer. Considered were whether and what potential confounding factors were accounted for and what steps were taken to avoid biases. We assessed how adequately results were presented, whether data were missing, and whether it was possible to obtain data from the report for calculations, if needed.p < 0.0001), with scores ranging from 10 to 25 for reviewer one, and 9 to 23 for the second reviewer. Continuous scores were classified into tiers  by dividing the range to create tertile cutpoints for the categories. Both authors re-assessed articles for which rater tier scores differed from each other. Agreement between authors in quality tier scores was reached after a second review and discussion.A 26-point quality assessment tool  was deveWe used a random effects model to calculate pooled, meta-effect estimates (as approximations of meta-RRs), which assumes that the study-specific effect sizes come from a random distribution of effect sizes according to a specific mean and variance . For thep-value < 0.10 from tests of homogeneity to suggest an investigation into potential sources of heterogeneity. As data permitted, we explored potential sources of variability by calculating meta-effect estimates for subgroups within the studies based on: geographic region where the study was based ; type of study design ; year of study publication ; sex ; race ; type of farmer ; farming duration ; type of brain cancer (glioma); source of exposure classification ; and quality assessment tier .We quantitatively assessed variability in pooled estimates across studies using a In sensitivity analyses, we calculated meta-effect estimates for studies that specifically reported data for a more highly exposed subcohort of workers and for farmers who had documented exposure to pesticides.We examined the relative influence of specific studies or combinations of studies by calculating pooled estimates with the removal of a study(ies) and comparing them to the overall estimate.We used the Preferred Reporting Items for Systematic reviews and Meta-Analyses (PRISMA) flow diagram to report identified records, excluded articles, and included studies . All anaOf the 61 total studies covered by the two previous meta-analyses, 16 were included in both reviews. Two articles ,31 from All but three of the articles in the current meta-analysis included study populations from North America or the European region, including the United Kingdom, Scandinavia, and Iceland. Twenty cohort studies, 20 case-control studies, and 12 PMR studies contributed data from predominantly white males. Farming occupations were assessed from registries or other administrative lists, death certificates, personal interviews, and self-reports. From the descriptions provided, the type of farmer could be determined for 31 (60%) studies, with 16 of these providing data on crop farming and 13 for livestock. Four studies reported data on duration of farming, 19 included data for one or more highly exposed subcohorts, and eight provided risk estimates specifically for glioma brain tumors. The majority (77%) of studies were judged to be of fair or poor quality .Forty of the 52 (77%) studies reported positive associations between farming and brain cancer, with effect estimates ranging in magnitude from 1.03 to 6.53. Of the 11 studies that reported negative associations, only three had estimates that were different than the null . These tp-homogeneity < 0.0001). The elevated risk was apparent and did not vary by the publication date of studies, either prior to 1990, between 1990 and 1999, or in the year 2000 or later, as indicated by overlapping confidence intervals of these subgroup risk estimates. The meta-effect estimate calculated for US studies  was comparable to the overall estimate. The risk of brain cancer from farming in Europe and the United Kingdom  or Canada  was not different than the null. Among the epidemiologic study designs, the pooled estimate from cohort studies did not indicate an elevated risk, whereas, in case-control studies, the meta-effect estimate was 1.39 . Results did not indicate a difference in the effects of farming on brain cancer between men and women or by the source with which farming exposure was classified. Farming among white populations was associated with a higher risk of brain cancer than among non-white populations , suggesting that farming is associated with a 13% increase in risk of brain cancer morbidity or mortality. However, the studies combined in this pooled analysis were significantly heterogeneous  was associated with a greater risk of brain cancer than crop farming . The risk of a glioma tumor type was not statistically significantly elevated . The four studies ,50,76 thNineteen studies included data for a more highly exposed subcohort; the meta-effect estimate  for these studies was similar in magnitude to the overall estimate and statistically significantly elevated. Ten studies provided documentation to substantiate that farmers had been exposed to pesticides. The pooled estimate for these studies was 1.22 .Despite the various subgroup analyses, significant heterogeneity remained in the majority (88%) of the pooled estimates.This comprehensive review and meta-analysis encompassing 42 years of the epidemiologic literature and updating two previous meta-analyses by 20 years supports an association between farming and brain cancer incidence and mortality. The magnitude of the overall meta-risk estimate for 52 studies is between that of the Khuder et al. 1998  meta-anaIn the Khuder et al. meta-analysis , the metIn the Acquavella et al. meta-analysis , the resMeta-analyses are a quantitative method to derive a more statistically precise risk estimate and a better understanding of the consistency (or inconsistency) of findings in the literature . AdditioFarmers may be occupationally exposed to pesticides by applying them to control agricultural pests, including weeds, and to eliminate rodents. Thousands of pesticides, including insecticides, herbicides, and fungicides, are widely used worldwide in agriculture . PesticiIn the current meta-analysis, we extended beyond previous work and calculated a meta-estimate for farmers exposed to pesticides. Restricting analyses to the ten studies in our meta-analysis that convincingly documented pesticide exposure among farmers suggested that pesticide use in farming occupations was associated with greater than a 20% elevated risk of brain cancer. The observation that farming–brain cancer associations varied by type of farmer could suggest that exposures, including those to pesticides, are different by crop or livestock farming. Many chemical classes of pesticides used in the US, for example, vary depending on the type of crop , and insecticides are commonly used externally and systematically in livestock to control parasites . DiffereIn addition to pesticides, farmers are also exposed to other potentially hazardous chemical and biological agents, such as solvents, fuels, and oils, biologically active dusts, viral and bacterial exposures from farm animals, nitrates from fertilizers, and other chemicals widely generated in agriculture, such as formaldehyde, ammonia, hydrogen sulfide, and phenols, all which can have adverse health effects in their own right ,83,84,85Gliomas, which arise from the supportive tissue in the brain called the glia, account for 80% of intracranial primary malignant tumors . Our subPublication bias, the propensity to publish studies with positive results but not publish studies with negative or null results, is a consideration in meta-analyses . Both KhIn summary, despite heterogeneity among studies, we conclude that the synthesis of evidence from over 40 years of epidemiologic literature supports an increased risk of brain cancer from farming with its potential for exposure to chemical pesticides. Increasing organic farming practices is one means to reduce the exposure of farmers to chemical pesticides.


Helicobactor pylori infection, these results were inconformity. Therefore, a comprehensive meta-analysis was performed to evaluate the TLRs gene polymorphism and susceptibility to H. pylori infection.There were many case-control studies performed the association between TLRs gene polymorphisms and the correlation of H. pylori infection, all the included data will be analyzed with the software of Review Manager 5.2 and STATA 14.2.Eligible studies were searched from PubMed, EMBASE, Web of science, Cochrane library, CNKI, CBM, Wan Fang Database and VIP Database, all the databases were searched from inception to December 2020. OR with the corresponding 95% CI were presented as associations between certain TLR gene polymorphism and the risk of H. pylori infection susceptibility.This study will provide a high-quality evidence to find the TLR gene polymorphisms with H. pylori infection.This study will explore which TLR genotype increase the risk of  So far, 10 members of TLR family have been found in human body, each member has the ability to recognize specific microbial pathogens and the initiation of innate immunity.,14 Recent years, many case-control studies performed the association between TLR gene polymorphisms and the susceptibility of H. pylori infection. However, limited statistical power to detect small-effect single nucleotide polymorphisms (SNPs) and the results were in conformity. To further evaluate the relationship between TLR gene polymorphisms and the risk of H. pylori infection, a meta-analysis was conducted in our present study.Toll-like receptors (TLRs) are cell transmembrane and pathogen-associated molecular pattern (PAMP) receptors that play an important role in the recognition of 2H. pylori infection susceptibility.The objective of this study was to comprehensively evaluate the TLR gene polymorphisms associated with 3 The protocol was registered on International Platform of Registered Systematic Review and Meta-analysis Protocols (INPLASY202130009).The authors conducted this systematic review and meta-analysis protocol according to the Preferred Reporting Items for Systematic Review and Meta-analysis Protocol Guidelines (PRISMA-P).3.13.1.1P value of Hardy–Weinberg equilibrium (HWE) test in the control group was greater than.05. For duplicate publications, family-based association studies, case reports or case series studies, no available genotype/allele frequency, abstracts, reviews, comments, letters and conference presentations were excluded.Case-control study design in human, peer reviewed publication. provide the available genotype/allele frequency between cases and controls, the 3.1.2H. pylori infection and were taken serum samples will be included in the meta-analysis. Healthy population served as controls. No restrictions were placed on age, gender, country.Participants affected by 3.1.3H. pylori infection risk comparisons.3.23.2.1We searched the following seven databases: PubMed, EMBASE, Web of science, Cochrane library, China National Knowledge Infrastructure (CNKI), China Biology Medicine (CBM), Wan Fang Database for Chinese Technical Periodicals and VIP Database, all the Database were searched from inception to December 2020.3.2.2The search strategy will use a combination of MESH terms and free terms. Details of the PubMed search strategy appear in Supplementary Materials 1.3.33.3.1The extraction work were independently by two investigators (LL and YT), for disagree evaluation, an agreement was reached following consultation and discussion, the data information of all studies included: name of the first author, year of publication, country, ethnicity, sample size/age of the cases and controls, and genotype/allele frequency between cases and controls, HWE were based on genotype frequency of certain TLR gene polymorphisms in control groups, no minimum number of samples defined for inclusion in our meta-analysis. The screening process was shown as Figure 3.3.2 We assessed it from the selection, the comparability, and the exposure in both cases and controls, the NOS score vary from zero to nine, in which (score 0–5) was low quality study and (score 6–9) was high quality study.The quality of included studies was assessing by the Newcastle–Ottawa scale (NOS).3.4H. pylori infection in the allele, codominant, dominant and recessive models, the significance was determined by Z-test, with a P value < .05 as significant level.All statistical analyses were performed by Review Manager 5.2 version software . OR with the corresponding 95% CI were presented as associations between certain TLR genetic polymorphism and the risk of 3.52 statistics, when P value of Q test was greater than 0.1 or I2 was less than 50%, fixed effects model was used for meta-analysis, otherwise, random effects model was carried out. If there were high heterogeneity, we will be found the potential sources of heterogeneity by using the methods of sensitivity analysis and subgroup analysis.The heterogeneity assumption was checked by Q-test and I3.6H. pylori infection by ethnicity or country of participants.We will conduct a subgroup analysis of the TLR gene polymorphisms most associated with 3.7Sensitivity analysis were performed by deletion of each single study in the meta-analysis to reflect the influence of the individual data-set on the pooled OR.3.8The publication bias will be evaluated by using the funnel plots combine with statistical test(Egger and Begg test), and the Egger and Begg test will be performed by STATA software with version 14.2 .4H. pylori infection may be related to the age, gender, race, genetic factors, geographical location and socio-economic status of the host.,18 TLR is the main component of innate immune system and its expressed in different cell types recognize H. pylori through different signal pathways to activate the immune system in vivo. In recent years, researchers have paid more attention to the influence of host genetic factors on H. pylori susceptibility.,21 Recent years, many case-control studies performed the association between TLR gene polymorphisms and the susceptibility of H. pylori infection, but there is no consensus in the literature. Therefore, it is necessary to conduct a systematic review to assess the TLR gene polymorphisms and the risk of H. pylori infection.Conceptualization: Weisheng Luo.Data curation: Liying Lu.Methodology: Yan Tang.Project administration: Meiwen Tang.Resources: Jianxiang Li.Writing – original draft: Xiaocong Ma.


SNCA are an established risk factor for PD and are candidates to modify disease presentation and progression. This systematic review aimed to summarize all available primary research reporting the association of SNCA polymorphisms with features of PD. We systematically searched PubMed and Web of Science, from inception to 1 June 2020, for studies evaluating the association of common SNCA variants with age at onset (AAO) or any clinical feature attributed to PD in patients with idiopathic PD. Fifty-eight studies were included in the review that investigated the association between SNCA polymorphisms and a broad range of outcomes, including motor and cognitive impairment, sleep disorders, mental health, hyposmia, or AAO. The most reproducible findings were with the REP1 polymorphism or rs356219 and an earlier AAO, but no clear associations were identified with an SNCA polymorphism and any individual clinical outcome. The results of this comprehensive summary suggest that, while there is evidence that genetic variance in the SNCA region may have a small impact on clinical outcomes in PD, the mechanisms underlying the association of SNCA polymorphisms with PD risk may not be a major factor driving clinical heterogeneity in PD.There is great heterogeneity in both the clinical presentation and rate of disease progression among patients with Parkinson’s disease (PD). This can pose prognostic difficulties in a clinical setting, and a greater understanding of the risk factors that contribute to modify disease course is of clear importance for optimizing patient care and clinical trial design. Genetic variants in  While PD is characterized by motor symptoms such as resting tremor, bradykinesia, rigidity, and postural instability, patients also experience a variety of non-motor symptoms, including cognitive impairment and dementia, sleep disorders, depression, anxiety, hallucinations, and hyposmia3. These non-motor symptoms complicate the clinical management of the disorder and are significant determinants of poor quality of life for patients and their caregivers. The observed heterogeneity can also pose prognostic difficulties in a clinical setting, and a greater understanding of the risk factors that contribute to modify disease course is of importance in terms of informing patients more accurately how they may be affected by the disease and developing personalized medicine strategies for patient management. Further, there is increased interest in targeting patients with a higher risk of specific disease outcomes in tailored clinical trials.There is great heterogeneity in both the clinical presentation and the rate of disease progression between patients with Parkinson’s disease (PD)SNCA, the gene encoding the α-synuclein protein that constitutes the major protein component of Lewy bodies (LBs)4. The abnormal aggregation of α-synuclein has a primary role in the formation of the LBs and other α-synuclein pathological aggregates and is regarded as a critical step in the molecular pathogenesis of PD5. In familial cases of PD, specific mutations or copy number variations (CNVs) of SNCA have been shown to cause increased production of α-synuclein and this is correlated with increasing disease severity7. Such mutations and CNVs of SNCA are rare in the general PD population, but candidate gene studies and genome-wide association studies (GWAS) have established the SNCA locus as a risk factor for idiopathic PD9 and accordingly as a promising candidate for disease modification.A logical first step in seeking genetic modifiers for progression in PD is to examine well-established PD susceptibility genes. These include SNCA region with clinical outcomes or age at PD onset has received much attention, but the use of different clinical scales, populations, and study designs complicates the comparison of the results, and no clear disease-modifying polymorphisms have been identified so far. For a better understanding of the impact of SNCA genetic polymorphisms on the progression of disease, a comprehensive overview of the associations between clinical evaluations and SNCA genotypes in PD is necessary. This systematic review aims to summarize and compare all available primary literature that has evaluated the association between SNCA polymorphisms and any PD clinical outcome. The review is limited to the analysis of patients with idiopathic PD, but given the broad symptomatology accredited to PD, no restrictions were placed on the clinical outcome measures evaluated in the studies. The potential contribution of SNCA polymorphisms to the clinical manifestations of PD is summarized. Further, the implications of the results for our understanding on the pathophysiology of PD and the design of future studies are discussed.The analysis of common polymorphisms in the n = 37) or Asian ancestry (n = 24). Only one study population was from South America (Brazil). All studies used defined diagnostic criteria for the diagnosis of PD  and variations in the complex microsatellite D4S3481 known as REP1, located approximately 10 kb upstream of the translational start of SNCA. The location and population frequency (European and East Asian) of each polymorphism is provided 16 and Asian (n = 2)17 ancestry, did not find evidence that the SNCA REP1-risk genotype impacts AAO in PD  Consortium, including some of the previously cited studies, found higher REP1 scores  to be associated with an earlier AAO (hazard ratio (HR) = 1.06; p = 0.01)21.Seven small studies, including patients of European 23. These data are consistent with reports from both a Chinese population of 685 cases 25 and a Spanish population of 898 cases 24, each of which reported that carriers of the rs356219-G allele had a significantly younger AAO than carriers of rs356219-A allele. Notably, the results from the Spanish population were extended to identify a three-loci epistatic combination of rs356219 with RPTOR rs11868112 and RPS6KA2 rs6456121 that was associated with AAO in both the discovery cohort  2.89; p < 0.0001) and a replication sample from the International Parkinson Disease Genomics Consortium . The most recent AAO study was also the largest study to be identified in this review, achieved by combining cases from the IPDGC with participants from the 23andMe research data set, which consisted of customers of the personal genetics company  who self-reported diagnosis and their age of PD diagnosis10: Blauwendraat et al. took two approaches, first performing a GWAS for AAO10. The top SNP associated with PD AAO in this GWAS was rs356203, resulting in an earlier AAO by approximately 0.6 years (p_meta = 1.90E−12). Interestingly, rs356203 has previously been shown to be associated with an increased risk of PD8 but has not been investigated for an association with other clinical outcomes in PD. Second, the authors analyzed an updated set of 44 PD-risk SNPs and identified variants in six different genes to be significantly associated with AAO, including the SNCA SNP rs365182 (p = 1.12 × 109)10.Five studies took a broader approach to identify genetic AAO modifiers by analyzing the association of SNPs across many genes previously associated with the risk of sporadic PD, including polymorphisms in the 31 or the REP1 polymorphism34  scores were found in two Asian populations for REP1–263 allele carriers 34 and rs11931074  −0.083; p = 0.035; UPDRS III, SC −0.140; p ≤ 0.001)28. However, these results were not replicated by other studies 36. Cooper et al. showed an association between rs356182 and tremor-dominant (TD) vs. postural-instability gait disorder motor subtypes (corrected p = 0.04), which they subsequently replicated in an independent multi-center cohort (p = 0.002). In both the discovery and replication cohorts, the authors found that the rs356182-GG genotype was associated with a more TD phenotype29. Cooper et al. built on these findings and showed that rs356182 also predicted slower progression of motor impairment, with the annual rate of increase in UPDRS III scores found to be approximately 1 point per year less in rs356182-GG carriers compared to those with the AG or AA genotypes (p = 0.01)29. Further evidence that SNCA polymorphisms predict the rate of motor symptom decline in PD was reported in work from the Parkinson Environment and Genes cohort, a prospective, population-based study of patients in California, USA. In this study, the risk of faster decline of motor function (defined by a change of >5 points per year in UPDRS III score) was increased fourfold in carriers of the REP1–263 allele 32.In cross-sectional analyses, significant associations of SNCA polymorphisms and the development of motor complications related to dopaminergic treatment (Supplementary Table p = 0.039), while no difference was observed in the development of LIDs between the two groups (unadjusted p = 0.907)26. Two studies assessed wearing off (WO): in an Italian study, carriers of the REP-263 allele were found to have a higher risk of WO compared to noncarriers at 10 years from disease onset , although this result was not significant in multivariate Cox regression analysis18. In a separate Chinese study, rs11931074 was not found to be associated with WO (p = 0.520) in 724 patients with a comparably short disease duration 28. Given the higher risk of motor problems with increasing disease duration, additional studies of more advanced PD will be needed to validate these results.Few studies have addressed a possible association between annual change in H&Y stage22, while in other studies these same SNPs were shown to be significantly associated with either a longer time to a 0.5-point increase in H&Y stage 37 or a longer time to reach H&Y stage ≥3 (mild-to-moderate disability) 31. Similarly, REP1 score was reported to be associated with a longer time to reach H&Y stage ≥4 (severe disability) 38, while reports on the association of the REP1 polymorphism with H&Y stage at the time of examination are conflicting36. Shi et al.  reported no associations with H&Y stage in an exploratory study of 9 genetic variants from 6 different genes that included REP136, while, in a smaller study (n = 172), Ng et al., found an association between REP1–263 and H&Y stage at examination 34. These studies also vary greatly in the rigor of their design, but variability in the application of the H&Y scale makes a direct comparison of the results particularly challenging.The Hoehn and Yahr (H&Y) scale provides an overall assessment of severity of PD based on clinical features and functional disability and, since its introduction, has remained the most widely used scale to describe the general severity of PD. Fourteen studies including a total of 31 different polymorphisms used the H&Y scale to measure PD severity , measured by the time to a 1-point decrease on the MoCA scale37. Conversely, a Brazilian study of 105 patients found that both rs356219 heterozygotes  and homozygotes  had significantly increased risk of cognitive impairment as defined by an education-corrected Mini-Mental State Examination (MMSE) cut-off35. Given Mata’s findings and the small number of participants and events observed in the latter two studies, there is currently limited support for a role of rs356219 in modifying cognitive impairment.Cognitive decline is an important aspect of PD as it brings a substantial additional burden for the patient, caregivers, and the healthcare system. In 8 of the 21 studies that assessed the association between 36 p = 0.01) in a Chinese population of 291 patients39. In contrast, studies of two European populations of patients with PD from the UK (CamPaIGN)42 and Spain43 found no link between rs356219 and the risk or development of dementia. All four studies used clinical diagnostic criteria for PDD  or via proxy (Alzheimer’s Disease Dementia Screening Interview)38. The group of patients with higher REP1 scores (linked with highest risk of PD) were shown to have reduced risk of developing cognitive impairment 38. However, this protective effect of REP1 on cognitive decline was not replicated by an Italian study of 426 PD patients which showed that carriers of the REP1–263 allele (similarly linked with PD risk) showed significantly increased risk of dementia compared to carriers of shorter alleles 18 or by a Singapore study which showed that REP1–263 allele carriers were associated with lower MMSE scores 34. While these later results are more aligned with the predicted effect of REP1 on disease outcomes, it is difficult to directly compare the results because of the scales used to asses cognition and the handling of the REP1 genotype are different in all three studies.One of the largest studies to explore the impact of n = 6) and Brazil (n = 1) looked at depression or anxiety and a total of 11 SNCA polymorphisms (Supplementary Table n = 105) Brazilian population, the rs2583988-TT genotype was found to significantly reduce the risk of depression, defined using the Beck Depression Inventory 35, and second, a large study in China (n = 1047) found that a REP1 “risk allele” (in this study defined by the copy number of a CA repeat) was associated with a decreased risk for the presence of mild-to-marked depression defined using the Hamilton Rating Scale for depression 44. Neither of these findings has been replicated.Anxiety and depression are two of the most common mental health symptoms that affect people with PD. Seven studies from China  in the multi-center NeuroGenetics Research Consortium45. In the second study, with patients recruited in Italy, REP1–263 allele carriers were shown to have a 2.69-fold higher risk of developing visual hallucinations over the course of disease (p = 0.001) recorded from retrospective review of clinical records18.As PD progresses, many patients will develop psychotic symptoms, primarily hallucinations, which is considered an important clinical milestone in the course of PD. However, only two studies reported the association between 46. In a study of 325 patients with PD from Norway47, possible RBD (pRBD) was assessed using the RBD screening questionnaire . pRBD was significantly associated with the SNCA SNP rs3756063  but not rs356165 or rs2245801. This result was replicated for rs3756063 in 382 patients from the Parkinson’s Progression Markers Initiative (PPMI) cohort with marginal significance 47. Recently, a large study analyzed the effect of four other SNCA SNPs, selected for their association with risk of PD or isolated RBD, in the Norwegian and PPMI cohorts and an additional cohort from McGill University48. Meta-analysis of the three cohorts identified a significant association of rs287004 with a reduced incidence of pRBD in PD 48.Sleep disorders, including excessive daytime sleepiness and rapid eye movement (REM) sleep behavior disorder (RBD), usually increase in frequency over the course of PD. RBD is a parasomnia strongly linked to synucleinopathies and characterized by the absence of muscle atonia during REM sleep50 diagnosed RBD using the “gold standard” set by the International Classification of Sleep Disorder-II criteria51. These studies did not identify an association of RBD with multiple SNCA genetic polymorphisms in Chinese49 or Italian50 populations (Supplementary Table n = 5750 and 15249), as typified by studies of clinically defined RBD using overnight polysomnography, larger studies will be needed to characterize the impact of SNCA polymorphisms in formally diagnosed RBD in PD.Two of the studies identified in this systematic review52. However, only three small studies identified in this review have explored the association of SNCA polymorphisms with hyposmia, each using the 16-item odor identification “Sniffin’ Sticks” (SS-16) tool to measure hyposmia in PD. Of these, 1 Chinese study of 50 patients followed for 30 months suggests that of the six SNPs selected, rs894278-G  and rs356219-G  were associated with slower olfactory impairment37. Another Chinese study of 218 patients report that the rs11931074-TT genotype increased the risk of hyposmia more than 3-fold compared with the GG genotype  but did not find any significant associations for rs89427853. The third study (n = 295) found no associations for SS-16 outcome measures and REP1 in a population from the Netherlands  identified in this review, no significant associations were reported with a clinical outcome or AAO, although it is notable that 31 of these SNPs were only included in 1 or 2 studies . Clinical outcomes were included in 17 of the studies analyzing rs356219 and 11 of the studies analyzing REP1. Multiple significant associations, spanning the spectrum of the clinical outcomes identified in this review, were reported for both rs356219 and REP1, but notably none were replicated in independent populations  but was also notable in the application of the UPDRS and MMSE  statement was used as a guide to ensure a rigorous review process; however, we note that the protocol for our review was not published before the search was initiated. The strengths of our review include the broad inclusion criteria, which captured studies of any clinical outcome associated with PD and including any SNCA polymorphism. Furthermore, several aspects of the design of each study were evaluated using pre-determined criteria in order to provide an assessment of the sources of bias in the studies.Several limitations must be considered in interpreting the results presented in this review. Despite the large number of studies identified, the heterogeneity in study design meant that it was not possible to compare the effect sizes among the studies or perform a meta-analysis. Further, we only reported the effects of multiple testing if the authors themselves included this as part of their analysis plans, although the assessment of potential sources of bias served to highlight some studies in which this was a weakness in the design  years and the mean age at examination was 63.7 (±4.5) years. This is typical of general research studies, which are generally unrepresentative of the population age distribution of PDSNCA SNPs in modifying disease course in PD and suggests that the genetics of PD risk and progression are largely separate. However, methodological aspects may explain some of the discrepancies in the effects of polymorphisms across the different studies. Most significant associations identified in this review found that SNCA polymorphisms had minor effects on the clinical outcomes and so, even if validated, these findings are not currently relevant in a clinical setting. However, given that α-synuclein is a major drug target, they have the potential to be used as a genetic risk stratification tool to reduce variability in clinical trials, and so further studies using large, well-designed cohorts are warranted.The lack of robust findings casts doubts on the relevance of 64. A systematic review of literature was performed to summarize all available primary research reporting the association of SNCA polymorphisms with clinical features of PD. The following inclusion criteria were used: subjects were patients with PD diagnosed according to defined diagnostic criteria ), genetic polymorphisms were within the SNCA gene region, outcomes were any clinical feature attributed to PD or AAO, the association of SNCA polymorphisms with outcomes was assessed statistically, and the source was primary literature in English. Full exclusion criteria are listed in Supplementary Table The review process was conducted according to the PRISMA statementSNCA AND (variant* OR polymorphism* OR SNP* OR GWAS OR genome-wide) (Supplementary Table n = 5) and from the wider PD literature (n = 3)  were retrieved, as available. An overview of the papers included in this review  and the following search terms: Parkinson* AND ture n =  (identifFurther information on research design is available in the Supplementary InformationSupplementary Data 1Reporting Summary


The serum Vitamin D status in patients with vitiligo is ambiguous when compared to controls. A systematic review and updated meta-analysis were conducted to evaluate the association between Vitamin D and vitiligo.Relevant studies were identified by searching PubMed and other databases. The random effects model was used to obtain standardized mean differences and pooled correlation coefficients. Meta-regression and sub-group analyses were conducted to explore heterogeneity. The presence of publication bias and the study robustness were tested using funnel plot and sensitivity analyses, respectively.This meta-analysis finally included 31 studies. Compared with controls, vitiligo patients showed significantly decreased serum Vitamin D levels . The sub-group analysis showed that vitiligo patients with indoor/urban work had a significantly lower Vitamin D level when compared to their outdoor/rural counterparts . The sensitivity analysis indicated that no single study had a significant influence on the overall outcome, suggesting the robustness of this meta-analysis.Varied sample sizes and heterogeneous study populations from different countries are the limitations of this study. However, the between-study heterogeneity has been addressed by the random-effects model with meta-regression and sensitivity analyses.This meta-analysis showed significantly decreased Vitamin D level in vitiligo, and its association with indoor/outdoor type of work of vitiligo patients. This study highlights the need to assess Vitamin D status for improving its level in vitiligo. Vitamin D (Vit-D) has pivotal involvements from bone structure to brain function and is also vital for healthy skin.4The available studies in the literature are inconsistent on Vit-D levels and Hypovitaminosis D is still controversial in vitiligo.https://www.crd.york.ac.uk/PROSPERO/display_record.php?RecordID=142156).The criteria of Preferred Reporting Items for Systematic reviews and Meta-Analysis (PRISMA) were followed in the conduction and reporting of this study. The study protocol has been registered with PROSPERO having registration number: CRD42019142156  OR "vitamin D" (All Fields) OR "ergocalciferols" (MeSH Terms) OR "ergocalciferols" (All Fields)) AND ("vitiligo" (MeSH Terms) OR "vitiligo" (All Fields)). The other terms used for Vit-D were 25-Hydroxyvitamin D, 25-Hydroxycholecalciferol, calcidiol, 1.25-dihydroxyvitamin D3, and calcitriol. Also, the Google scholar, Cochrane library, Springer’s author mapper, science direct, clinical trial registries, and CINHAL databases were also searched for articles that reported circulatory Vit-D levels in vitiligo patients. Furthermore, bibliographies of published articles were manually hand-searched to identify additional studies. Two authors have independently performed the literature search, and any discrepancies were resolved upon discussion. Efforts have been made to obtain data, if any, from unpublished sources. When required, the corresponding authors of respective articles were contacted through e-mail to obtain data/clarification. All the literature searches were carried out prior to 5The following criteria are followed for individual studies to be included in this meta-analysis: (1) papers published in a peer-reviewed journal; (2) provided original data; (3) study subjects were human with a description of vitiligo and controls; (4) studies where serum Vit-D levels in vitiligo patients were compared with controls; (5) provided method description for Vit-D estimation; (6) articles written in English/Chinese with clear data; (7) studies diagnosing vitiligo clinically by a dermatologist. The exclusion criteria were as follows: (1) studies with no control group; (2) reported on other skin diseases; (3) Vit-D levels were only shown by pictograms with no data; (4) reviews, commentaries, and letter to the editor article types. If any duplicate articles were found, only a recent report or all relevant information from either of the studies was included.The following information has been extracted from each of the included studies; first author names, country, study type and year of publication, number of vitiligo and control subjects, male and female subjects in each group, mean/median values of age, disease duration, means and SD metrics of Vit-D, specimen used for its measurement, method and units, vitiligo type and history, disease activity, skin types, autoimmune comorbidities, and other study characteristics. A quality score evaluation of included studies was done according to the Newcastle-Ottawa Scale (NOS).Meta-analyses and sub-group analyses were conducted if three or more studies were reporting the serum Vit-D levels in vitiligo and control groups. We calculated the Standardized Mean Difference (SMD) and its 95% Confidence Interval (CI) as a summary statistic for the difference of Vit-D level between vitiligo patients and controls. Furthermore, we have also performed a meta-analysis of correlations between Vit-D and other parameters in vitiligo. The effect size for SMD and pooled correlation coefficient values were presented as a Z-score. The Z-score with a p-value of <0.05 was considered statistically significant.2. A p-value of <0.05 or I2 statistic of >50% indicated significant heterogeneity. A random-effects model was used to compute SMD. The meta-regression analysis was performed to identify the potential source of heterogeneity using several covariates such as age, sample size, number of males/females, disease duration, Vit-D measurement method, country, and year of publication. The risk of publication bias was studied by funnel plot asymmetry with Begg’s and Egger’s tests. However, to avoid the risk of bias, general search terms were chosen for a guaranteed retrieval for the inclusion of articles reporting Vit-D levels as a secondary outcome. In the case of a p < 0.05 suggesting a statistical significance, the “trim-and-fill method” was used to correct that bias. To test the robustness of this meta-analysis, a one-study leave-out sensitivity analysis was performed. All comparisons were two-tailed, and all analyses were conducted using the Review Manager Software version 5.3 which presents SMD as Hedges g; the difference between the two means divided by the pooled standard deviation, with a correction for sample bias. The MedCalc version 16.2.0 was used for the meta-analysis of correlations. The meta-regression and publication bias were respectively performed using OpenMeta and Comprehensive meta-analysis, version 3 software.The between-study heterogeneity was examined by Cochrane’s Q statistic and expressed as the percentages of I,A total of 167 papers were identified for Vit-D levels in vitiligo, of which 63 were retrieved for abstract and full-text review. Of these, a total of 31 articles did not deal with the circulatory Vit-D levels in vitiligo, therefore are excluded. In one more paper, no data has been reported and therefore is not included.272 = 95%; p < 0.00001), the pooled SMD (95% CI) obtained with random-effects model was −1.03 (−1.35 to −0.71). This result suggests a significant difference in mean concentrations of Vit-D between vitiligo patients and controls, and the overall effect size for the obtained mean difference corrected for standard deviations and sample bias was statistically significant .As shown in The results of subgroup meta-analysis based on country, Vit-D measurement method, vitiligo type, gender, skin type, vitiligo history, autoimmune comorbidities/history, disease activity (progressive/active or stable), job status (indoor or outdoor), and journal indexing (PubMed/Non-PubMed) were presented in The funnel plot analysis with Begg’s correlation (p < 0.001) and Egger’s regression tests (p < 0.001) indicate there might be a publication bias among the included studies . HoweverAvailable literature shows contrasting results with regard to Vit-D levels in vitiligo.The active form of Vit-D and Vitamin D Receptor (VDR) mediates several important biological functions such as calcium and phosphate metabolism, bone formation, growth, and differentiation. Studies have indicated that VDR polymorphisms are associated with Vit-D levels and exhibit a genetic predisposition for vitiligo.Our sub-group analysis produced interesting results. As there were ten studies from Egypt, maximum for any other country, we performed subgroup meta-analysis based on Egypt and non-Egypt studies 21 studies). In both cases, statistical significance was retained for a decreased Vit-D level in vitiligo. However, when non-Egypt countries were further stratified, the subgroup India (7 studies) showed a statistical significance, whereas the sub-groups of studies from China 3 studies) and Turkey (4 studies) did not produce a statistically significant difference in Vit-D levels between their respective vitiligo and control groups . This is1 studies studies It is noteworthy that Vit-D levels were significantly lower in indoor/urban vitiligo patients when compared to their outdoor/rural counterparts . This clWhen a separate meta-analysis was performed based on ELISA (15 studies) and non-ELISA studies (9 studies), the statistical significance was only retained in a group of studies using ELISA, while the non-ELISA methodology (chemiluminescence in particular) is an important factor contributing to the heterogeneity among studies. However, in meta-regression (24 studies), the ‘method’ used as a covariate did not show a statistically significant value. Considering the PubMed/MEDLINE indexed papers are of good quality with credible information, we repeated a separate meta-analysis on PubMed (18 reports) and non-PubMed studies (13 reports), and results retained statistical significance in either case .Considering the increased risk of vitiligo and the risk of autoimmune comorbidities associated with decreased Vit-D status, our meta-analysis and previous studies point towards the need for maintaining sufficient Vit-D levels by phototherapy with NB-UVB and/or Vit-D supplementation.Our study limitations include the between-study heterogeneity which could be attributable to heterogeneous study populations from different countries, different sample sizes, and differences in methodologies, varied environmental, genetic, dietary factors, and lifestyle habits, and other inherent limitations of case-control/observational study designs. However, as heterogeneity could not be completely eliminated in a meta-analysis, this has been addressed by applying the random-effects model for estimating the overall effect-size and by a cumulative and one-study leave out a sensitivity analysis to study whether any individual study would significantly affect the obtained overall effect-size. A one-study-removed sensitivity analysis revealed that no single study/observation had significantly influenced the pooled overall effect size. Therefore, our results could be considered robust. Further, all the included studies were matched for confounders like age and sex.In conclusion, our meta-analysis indicates decreased Vit-D status in vitiligo patients as compared to controls. These results of our meta-analysis highlight the need to assess Vit-D status for improving its level in vitiligo. Further studies on the role of Vit-D with vitiligo pathogenesis and severity are needed in the future.None declared.Seshadri Reddy Varikasuvu: Designed the study; conducted literature search; performed analysis; wrote the manuscript; read and approved the final manuscript.Sowjanya Aloori: Participated in literature search; selection criteria; data collection and revisions; read and approved the final manuscript.Saurabh Varshney: Participated in literature search; selection criteria; data collection and revisions; read and approved the final manuscript.Aparna Varma Bhongir: Assisted in literature search and revised the manuscript; read and approved the final manuscript.None declared.