Physical activity (PA) is considered a favorable preventive intervention for postpartum depression (PPD), but evidence defining a corresponding dose-response relationship is lacking. This meta-analysis was conducted to assess the protective effects of PA on PPD and define a potential dose-response relationship between them.PubMed, Medline, Embase, and Web of Science were searched from 1968 to May 2022. Only randomized control trials (RCTs) and prospective studies were considered, and the PICOS tool was used to identify eligible articles based on the inclusion and exclusion criteria. Effect-size estimates were unified as odds ratio (OR) and 95% confidence interval (CI). We calculated the ORs and their 95% CI for studies that did not report them using the Practical Meta-Analysis Effect Size Calculator.P < 0.001). Subgroup analyses indicated that studies conducted in Europe demonstrated a significant correlation between PA and reduced PPD risk . Concerning PA type, sports activity was associated with relieving PPD symptoms , while work  and household activities  contributed to a greater risk of PPD. Our dose-response analysis revealed a reverse J-shaped trend between ascending PA duration and PPD incidence.A total of 23 studies were eligible, including 14 RCTs and 9 prospective cohort studies. The overall analysis showed a statistically significant positive association between PA and PPD prevention (adjusted OR = 0.73; 95% CI: 0.61–0.87; This meta-analysis identified PA as a potential intervention to reduce the risk of PPD. The dose-response analysis revealed that at least 90 min of PA per week could efficiently decrease the risk of PPD.https://www.crd.york.ac.uk/PROSPERO/, identifier: CRD42022335731. Postpartum depression (PPD) is a specific type of mental disorder that might occur up to 1 year after giving birth and affects 5–25% of new mothers –5. In sevia breastfeeding, and the other three therapies are usually expensive, time-consuming, and require a lot of energy that might not be at the disposal of the suffering mother. Thus, newer and more efficient primary PPD prevention strategies are required to improve the recovery process of these patients.At present, among the many interventions recommended for treating PPD are antidepressant medications, cognitive behavioral therapy (CBT), psychosocial support through support groups, electroconvulsive therapies, and complementary therapy , 12. HowNigella sativa, a kind of herb, is helpful in the treatment of asthma, liver and kidney disease, influenza, and gastrointestinal problems  recommends 20–30 min of daily moderate-intensity PA to stay healthy during pregnancy (The three main categories of PA during pregnancy include (1) physical exercise, a subset of planned, structured, and repetitive PA, such as aerobic exercise, stretching and breathing exercises, walking programs, strength exercises, pilates, and yoga; (2) housework PA, such as child care and household chores; and (3) work activities. delivery . In addiregnancy . In 2014regnancy , 22. In regnancy . Thus, tIn this study, we conducted a meta-analysis to determine the effects of PA on PPD and to define the dose-response relationship between them to provide more detailed evidence for health authorities and pregnant women to reduce the risk of PPD.This meta-analysis was conducted following the guidelines of the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) statement  and the A literature search was performed from 1968 to May 2022 on PubMed, Medline, Embase, and Web of Science databases. The PICOS principle of this meta-analysis was as follows: (P) Population involving women who were pregnant or up to 1 year postpartum; (I) Intervention involving all kinds of physical activities; (C) Comparator including standard prenatal and postpartum care; (O) Outcomes dicussing the effect of PA on PPD symptoms; and (S) Study type including randomized controlled trials (RCTs) and prospective cohort studies. The detailed search terms are presented in the via a structured clinical interview and a validated tool  or judgEndnote version 20 was used for literature management to file the literature search records. The study selection process, conducted by two authors (M. Y. and H. C.), was performed following these three steps: First, two authors screened the titles of all the recovered articles and removed duplicates. An article was included if at least one reviewer judged it qualified. If there were any doubts about the studies, they were included in the abstract review phase. Second, the selected studies were filtrated by reviewing their abstracts and reevaluated by the two independent reviewers according to the inclusion and exclusion criteria. Finally, using standardized eligibility criteria, the two independent reviewers assessed the full texts of potentially relevant studies. If the reviewers disagreed on any of the three steps, a third independent reviewer from our group intervened to make a judgment after mutual discussions.The two reviewers (M. Y. and H. C) separately extracted the following data from the eligible studies: first author, year of publication, the country where the study was conducted, gender, sample size, study type, age of the population, PA type, frequency and duration of PA, PA intensity, and assessment of PPD.The two authors independently evaluated all eligible studies for bias using the Risk of Bias in Non-Randomized Studies of Interventions (ROBINS-I) assessment tool , which cP-value of < 0.05 was considered significantly different. The effect size of this meta-analysis was odds ratio (OR). For the studies that did not report ORs, we calculated their ORs and corresponding 95% confidence intervals (95% CI) using the Practical Meta-Analysis Effect Size Calculator  was used for data management and analysis. A two-sided lculator . Generallculator , 30, whiI2) statistic was used to quantify potential heterogeneity among the studies. Obvious heterogeneity was determined for I2 ≥50%. Publication bias was assessed using Begg's funnel plots and Egger regression asymmetry tests. The trim-and-fill method was applied to reckon the number of missing studies in theory. Sensitivity analysis was performed to judge the solidity of the overall results by consecutively omitting the study.The inconsistency index  met the inclusion criteria of this study, comprising a total of 186,412 subjects. The detailed selection process is presented in The baseline characteristics of the 23 , 31–49 eThe items of the ROB assessment of each study are listed in P < 0.001)  .P = 0.014), and no potentially missing study was shown by a further filled funnel plot.Publication bias was assessed using Begg's funnel plot and Egger's test. The results showed that Begg's funnel plots were not perfectly symmetrical , Egger'sA sensitivity analysis was performed to assess the influence of individual studies on the pooled ORs. By omitting one study each time, the results revealed that the corresponding pooled ORs did not change, which confirmed the reliability of our results.Our dose-response analysis of the relationship between PA duration and the risk of PPD demonstrated an inverted J-shaped trend . The resSubgroup analyses were performed to clarify the cause of between-study heterogeneity . SignifiP = 0.004), whereas studies conducted in America  and Asia  displayed a negative association between PA and decreasing PPD risk, with a statistical significance observed in the Asian subgroup.According to region, our results showed that studies conducted in Europe demonstrated a significant correlation between PA and reduced PPD risk , although the result was not statistically significant. In terms of gestation age of the pregnant participants, an obvious decrease in trend was observed for women in their second trimester (13–27 weeks)  and third trimester (28–42 weeks) .According to the age of pregnant women, women older than 30 years demonstrated a downward trend of PPD . However, work  and household activities  were identified as risk factors of PPD.Next, we divided PA into three major categories, namely, sports, work, and household activities. All kinds of physical exercises, such as yoga, walking, ball games, pilates, and other aerobics workouts, were included in sports activities. Work activities are the amount of muscle- and energy-demanding activities at work. All kinds of housework and childcare were included in household activities. According to our analysis of PA type, sports activities were associated with relieving PPD symptoms  or after pregnancy  demonstrated a downward trend for PPD risk. Study type analysis showed that the results of the RCTs demonstrated a statistically significant association between PA and reduction of PPD risk . In contrast, cohort studies suggested that PA could be regarded as a risk factor for PPD , though the result was not statistically significant.According to PA timing, PA during pregnancy  , which iPreviously, Nakamura and Waerden conducted a meta-analysis of 17 articles, including 6 RCTs and 11 observation studies, to assess the effects of PA during pregnancy on PPD . They reThere are some limitations to this study. First, only English-language articles were analyzed, possibly leading to a certain level of publication bias. Second, the error of dose-response analysis persisted and was inevitable in secondary analysis; however, the overall reverse J-shaped trend deserves further attention in the association of PA and the risk of PPD. Third, significant heterogeneity was observed in some subgroups, restricting the explanation of pooled effect-size estimates. Finally, since most of the original articles did not report detailed statistics on PA intensity, the relationship between PA intensity and PPD could not be explored in this meta-analysis, and further studies are still needed to fill this gap.In summary, PA was identified as a potentially beneficial intervention to reduce the risk of PPD, even though no statistical significance was observed in this meta-analysis. Dose-response analysis between PA and the risk of PPD showed a reduced risk of PPD with a longer PA duration. Furthermore, 90 min of PA per week could efficiently reduce the risk of PPD.The original contributions presented in the study are included in the article/Conceived and designed the experiments: DW, JL, and XB. Performed the experiments: MY, HC, DC, and CZ. Analyzed the data: MY, FL, XB, and HC. Contributed materials and analysis tools: MY, CZ, DC, FL, and YN. Wrote and revised the article: MY, JL, and DW. All authors read and approved the final manuscript before submission.This work was funded by the Clinical Study on Prevention of Recurrence of Uterine Fibroids After Minimally Invasive Surgery by Integrated Traditional Chinese and Western Medicine (2022-NHLHCRF-LX-02-0116).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.


HLA‐G and soluble HLA‐G (sHLA‐G) expression with gynecological cancers (GCs).Human leukocyte antigen‐G (HLA‐G) is implicated in several cancers and is considered to be an immune checkpoint regulator. We determined the association between polymorphisms in the 3′ untranslated region of HLA‐G14‐bp insertion (I)/deletion (D) and +3142C/G polymorphism in GC and to evaluate sHLA‐G expressionA meta‐analysis was conducted to examine the association between +3142C/G polymorphism and invasive cervical cancer (ICC) based on the allelic model G versus C , dominant GG+GC versus CC , and codominant GG versus CC  models, suggesting that the G allele and GG genotype are protective against ICC. In gynecological precancerous patients with human papillomavirus (HPV) infection, we found that the 14‐bp I/D under the codominant DD versus DI model  was of borderline significance. Soluble HLA‐G levels were significantly higher in patients compared with healthy controls . Stratification by cancer type revealed that the sHLA‐G levels were significantly increased in cervical cancer  and in subjects of Asian ethnicity .We revealed a significant association between the HLA‐G14‐bp I/D and +3142 C/G polymorphisms are associated with GC and HPV‐associated cervical cancer. In addition, we found significantly increased sHLA‐G levels in cancer patients. These results provide a basis for further studies in diagnostics and immunotherapy of GC. Human leukocyte antigen‐G (HLA‐G)14‐bp insertion (I)/deletion (D) and+3142 C/G polymorphisms are associated with gynecological cancer (GC) and human papillomavirus‐associated cervical cancer. Soluble HLA‐G is significantly increased in GC. Although aberrant HLA‐G expression has been reported to be associated with advanced tumor stage, metastasis status, and poor disease outcome, some discrepancies remain in various cancer types.Genetic variations in the The current evidence supports that HLA‐G is involved in cancer; however, discrepancies related to cancer heterogeneity and differences in methods remain. This study aimed to further investigate the role of HLA‐G in gynecological pathologies by analyzing data from existing published reports.22.1HLA‐G polymorphisms and gynecological cancers in the MEDLINE, Embase, and Cochrane databases (up to December 2021) using Medical Subject Headings and keyword combinations, including “HLA‐G,” “polymorphism,” “gynecological cancer,” and  “neoplasm”. Similarly, we searched for published studies involving the association between sHLA‐G levels and gynecological cancers by using “sHLA‐G,” “level,” “neoplasm,” and “gynecological cancers” as keyword combinations. Furthermore, we identified by a manual search and reviewed additional studies that were not indexed in the MEDLINE, Embase, and Cochrane databases. We considered that studies are eligible if they respond to the following inclusion criteriaHLA‐G polymorphisms in patients with gynecological cancers and in healthy controls andWe searched for published studies involving an association between We excluded studies that (1) included irrelevant, incomplete, and redundant data or (2) were systematic reviews, meta‐analyses, or case reports. For each included study, we extracted the following information: first author, publication year, country, cohort ethnicity, allele, and genetic frequencies of the studied HLA‐G gene polymorphisms, and mean and SD values of sHLA‐G. Two independent reviewers (K. T. and I. Z.) evaluated original studies and extracted the needed data. The meta‐analysis was conducted according to PRISMA guidelines.2.2HLA‐G gene polymorphisms (14‐bp I/D and +3142 C/G). In case of dichotomous data, the odds ratios (ORs) and 95% confidence intervals (CIs) were calculated. In case of continuous data, standardized mean differences (SMDs) were presented. The heterogeneity and variance between studies were evaluated using the I2 and Tau2 (τ2). The I2 values were interpreted according to the Cochrane guidelines.I2, which varied from 0% to 100%, and reflecting the proportion of variation among studies. I2 values of 25%, 50%, and 75% were considered low, moderate, and high, respectively. The random‐effects model, testing both sampling errors within the study and variances among studies, is used when heterogeneity among studies is significant (I2 > 50%).τ2 test measures the variance of the true effect sizes and may be used to test the variance of the effect size parameters across studies.pEgger value was reported. For the Begg and Mazumdar rank correlation test, we calculated a two‐tailed p‐value without continuity correction. The funnel plot is used to measure the study size.p < 0.05.We conducted a comprehensive meta‐analysis to test the allelic, recessive, dominant, homozygous, and codominant models for 33.1HLA‐G polymorphisms in gynecological cancers. Based on the title and abstract, we selected 42 studies for full‐text analysis. Some studies were excluded because they were reviews or meta‐analysesWe identified 83 eligible studies concerning For sHLA‐G and gynecological cancers, 101 studies were identified. Of these, 31 were selected based on the title and abstract for full‐text analysis. Studies were excluded as they were reviews3.2+3142 C/G polymorphism and invasive cervical cancer (ICC) under the allelic model, G versus C  , and codominant GG versus CC  models, indicating that theG allele andGG genotype may represent a protective allele and genotype against ICC. The +3142 C/G polymorphism did not exhibit an association with low‐grade squamous intraepithelial lesions (LSIL) or high‐grade squamous intraepithelial lesions (HSIL). Both conditions were classified as squamous intraepithelial lesions of the uterine cervix (Table I2 > 50%). In the HSIL group, only two studies were included, and therefore, heterogeneity was not detected (I2 = 0%). The variance observed in the forest plot reflects a difference in the true effect sizes rather than sampling errors, suggesting calculations in the random‐effects model. Egger and Begg'sp‐values were not significant , suggesting the absence of the sample size effect for all genetic models of ICC. A funnel plot was symmetric, revealing the absence of publication bias  Figure , dominans Figure . These r14‐bp I/D did not show a significant association in the study models   Figure . Howevert Figure .3.3p = 0.005) , more larger cohorts are required to confirm the conclusion of the present study. All calculations were performed with the random‐effects model, because heterogeneity was high (I2 > 75%). The high heterogeneity detected among the studies in the overall analysis (I2 = 96.5%) could be explained by the heterogeneity of cancers in addition to differences in the sHLA‐G detection methods  Figure . Stratifds Table .4HLA‐G gene polymorphisms (14‐bp I/D and +3142 C/G) and sHLA‐G with gynecological pathologies. By pooling several eligible studies, the meta‐analysis provided accurate data by increasing the statistical power and analysis resolution. In the present study, we evaluated eligible studies that included 1243 patients for HLA‐G 14‐bp I/D, 1240 patients and HLA‐G +3142 C/G, along with 4069 controls. Our data revealed a link between ICC and the +3142 C/G polymorphism. However, squamous intraepithelial lesions classified as precancerous lesions did not exhibit an association. We demonstrated that the G allele and GG genotype are likely protective against ICC under different contrast models . The presence of the G allele was proposed to enhance the affinity of HLA‐G mRNA for microRNAs, thus increasing the instability of the HLA‐G transcripts,G alleles may exhibit low HLA‐G expression and reduced immune tolerance associated with the HLA‐G molecule.Gynecological pathologies, particularly gynecological cancers, are multifactorial disorders in which genetic factors constitute strong determinants. HLA‐G, a potent inhibitory checkpoint of immune cell function, is highly expressed in gynecological cancers. The membranous HLA‐G is significantly expressed in cancers and plays a significant role in their diagnosis and progression.HLA‐G 14‐bp I/D did not reveal an association with precancerous lesions or with ICC  and, particularly, in cervical cancer . After stratification by ethnicity, the association of increased sHLA‐G expression with gynecological cancers was maintained only for the Asian population (p = 0.030). Our findings support sHLA‐G as a key biomarker for gynecological cancers. It plays an important role as a powerful immune regulator implicated in immune tolerance and the inhibition of immune cell function.The current meta‐analysis conducted on 299 cases and 375 controls revealed increased levels of sHLA‐G in all gynecological cancers  a significant association of HLA‐G +3142 C/G with reduced susceptibility to cervical pathologies and (ii) high sHLA‐G levels in patients with gynecological cancers, which support an important role for HLA‐G polymorphisms and sHLA‐G expression in the pathogenesis of gynecological cancers.This meta‐analysis of the association between Data collection, Data analysis and interpretation, Drafting of manuscript: Kalthoum Tizaoui. Medical validation, Critical data interpretation, Critical revision: Nadia Boujelbene. Critical data interpretation, Critical revision: Maroua Jalouli, Abdel Halim Harrath, Hadda‐Imene Ouzari, and Roberta Rizzo.Study conception and design, supervision, Data analysis, Drafting of manuscript, Critical revision: Inès Zidi.The authors declare no conflict of interest.


To clarify the association between smoking and stricture recurrence after urethroplasty.Pubmed, Web of Science, Embase, and Cochrane databases were searched with keywords: “urethroplasty,” “buccal mucosa graft urethroplasty,” “oral mucosa graft urethroplasty,” “excision and primary anastomosis urethroplasty,” “urethral stricture recurrence” until July 1, 2022. Inclusion and exclusion criteria were based on PICOS principles. The quality of included studies was assessed by Newcastle-Ottawa Scale (N.O.S.) system. Hazard ratio (H.R.), odds ratio (OR), and relative risk (RR) with 95% confidence interval (CI) were extracted or re-calculated from included studies. Meta-analysis was performed with Stata 15.0 based on univariate and multivariate data separately. Sensitivity analysis was performed to test the stability of the meta-analysis. I2 was calculated to evaluate heterogeneity. Publication biases were assessed by Egger's and Begg's tests. Funnel plots of univariate analysis and multivariate analysis were also offered.Twenty one studies with 6791 patients were involved in this meta-analysis. The analysis results of the two stages were consistent. In the univariate meta-analysis stage, 18 studies with 5811 patients were pooled, and the result indicated that smoking might promote stricture recurrence . Based on the adjusted estimate, 11 studies with 3176 patients were pooled in the multivariate meta-analysis stage, and the result indicated that smoking might promote stricture recurrence . There was no significant heterogeneity in both the univariate and multivariate stages.Our study demonstrates that smoking may prompt stricture recurrence after the urethroplasty. Quitting smoking may be a good option for patients undergoing urethroplasty surgery. When urologists deal with the urethral stricture patients by urethroplasty, one of the most worrying situations is the stricture recurrence . CurrentMany studies have pointed out that the use of tobacco, whether it is the inhalation of cigarettes or the use of tobacco powder or e-cigarettes, will increase the level of inflammation in the body. The increase in the level of inflammation in the body is closely related to the formation and aggravation of scars, The increase in the level of inflammation in the body is closely related to the formation and exacerbation of scars, which may further increase the possibility of recurrence of stricture after urethroplasty . In 2010It is worth noting that the statistical analysis of many retrospective studies of risk factors is usually divided into two parts, namely univariable factor regression analysis and subsequent multivariable regression analysis. Suppose only meta-analysis is performed on the results of multivariate regression analysis. In that case, it may lead to obvious selection bias (many studies only include variables that are significant in univariate regression in multivariate regression). Therefore, the results of single factor regression and multivariate regression analysis were combined separately in this study to get a more comprehensive result.The literature collection, data extraction, merging, and subgroup analysis methods used in this study are similar to those of our previous published studies .This meta-analysis was performed according to the principle of preferred reporting items for systematic reviews and meta-analysis (PRISMA) . This meDetailed inclusion criteria were as follows: Patients were treated with onlay with buccal mucosa or penile fasciocutaneous flap, oral mucosa, or any other type of substitution urethroplasty anastomotic urethroplasty or any combined urethroplasty techniques for anterior or posterior urethral strictures. Odds ratio (OR), hazard ratio (HR), and relative risk (RR) with a 95% confidence interval (CI) of risk factors should be offered or could be calculated. Only English-written studies were included. Exclusion criteria: Reviews, meta-analyses, letters, comments, case serials, and conference abstracts were excluded. Studies focused on hypospadias and pediatric patients and published earlier than 2000 were excluded. Studies that did not contain regression information or enough data to be used for secondary analysis were excluded. Since few studies offered detailed smoking history information such as tobacco type , smoking time, or whether current smoking, was smoking was not explicitly defined in this study. Because many studies based on adult cases did not strictly distinguish the cause of urethral stricture, this study did not exclude articles according to the cause of urethral stricture. The study was included in the analysis when the original study's smoking history variable was present. Two independent authors carried out all the title screening, abstract screening, and full-text review.All included studies were evaluated by Newcastle-Ottawa Scale (NOS) system, and two independent reviewers performed the evaluation procedure. Disagreements between the two authors should be determined by the third author (TJ). Accordi2. When I2 was larger than 50%, heterogeneity could be significant. If significant heterogeneity was detected, a random effect model should be applied. The primary outcomes were displayed with a forest plot. Subgroup analysis was also performed to get more detailed information. The variables included in subgroup analysis mainly include the research area, the number of participants, the recurrence rate of stricture, the type of estimates, the location of the stricture, and the type of urethroplasty surgery.Based on univariate and multivariate analysis results in this study, the relationship between smoking and stricture recurrence was pooled in a meta-analysis. All analysis was powered by Stata 15.0 software . Statistical significance was defined as P<0.05 in this study. Pooled estimates larger than 1 indicated that smoking would make patients more vulnerable to stricture recurrence. Heterogeneity was evaluated by IFurthermore, sensitivity analysis was performed to test the stability of meta-analysis results, and publication bias was tested by Egger's and Begg's tests. Funnel plots were used for publication-bias visual identification. After data synthesis, the final effect size should be the relative risk. This is a meta-analysis that tried to combine regression estimates. In many retrospective studies, only significant factors in the univariate logistic or Cox regression would be included in the multivariate logistic or Cox regression. Combining multivariate analysis results will undoubtedly bring selection bias if we only combine multivariate analysis results. It is necessary to perform a meta-analysis based on univariate analysis results.The previously published preprint has detailed and described all the analysis procedures .One thousand three hundred twenty-nine studies were identified from databases in total. 14 of included studies were carried out in the North American region, and 7 were carried out in other regions. Thirteen studies were carried out in recent 5 years, and no study earlier than 2000. Four studies focused on the anterior urethra, 5 on the bulbar urethra, 2 on the posterior urethra, and 7 studies did not limit stricture locations. Almost all studies did not define the cause of stricture. 8 of included studies used techniques for substitution urethroplasty with different materials, 3 of included studies focused on anastomotic urethroplasty. After duplicate removal, abstract screening, and full-text reading, 21 studies were finally involved in this meta-analysis. The detailed screening procedure is displayed in 2=0.0%, p=0.792)  with no significant heterogeneity found (Ip=0.792) . No signp=0.792) . Sensitip=0.792) . Further2=37.4%, p=0.09)  with no significant heterogeneity found (I p=0.09) . No sign p=0.09) . Sensiti p=0.09) . FurtherUrethral stricture is a kind of pathological stricture of the urethra, which can limit fluid transportation. Since the male urethra is significantly longer than the female urethra, and the posterior urethra is hidden in the pelvis, the urethral stricture can always bring many troubles to patients and urologists. Urethral stricture is a common urinary disease for males. There are 229-627 cases in every 100000 people, and in some susceptible groups, such as elderly men, the prevalence rate is as high as 0.6% . As one This meta-analysis revealed that tobacco consumption could increase the chance of stricture recurrence based on univariate and multivariate analyses. In the multivariate analysis stage, the sensitivity analysis result was not exactly stable, indicating that more multivariate analysis studies and adjusted estimates between smoking and stricture recurrence were required. In the univariate subgroup analysis, we found that anterior urethral stricture is most likely to be affected by smoking, increasing the risk of recurrence of the stricture. Similarly, patients who use substitution urethroplasty are more likely to be affected by smoking. However, similar effects were not found in the subgroup analysis based on the results of multivariable regression. This may be since fewer original studies provide the results of multivariate regression analysis. It was worth noting that in the subgroup analysis, we found that the pooled results based on HR always showed statistical significance, which may mean that the influence of smoking on the recurrence might have time-dependent distribution . However, since the primary studies did not provide a KM curve, we could not perform further analysis.Many studies have pointed out that smoking has a negative effect on healing wounds that cannot be ignored. Smoking temporarily reduces tissue oxygenation and aerobic metabolism. The inflammatory healing response is weakened by reducing inflammatory cell chemotaxis, migration function, and oxidative sterilization mechanism. In addition, the release of proteolytic enzymes and inhibitors would be unbalanced when the tissue was hypoxic or inflammatory substances were present. In addition to down-regulating collagen synthesis and deposition, reduced fibroblast migration and proliferation can also impair the proliferative response. A wound that delays healing would inevitably lead to repeated chronic local inflammation and tissue remodeling, which may be an important reason for stricture recurrence.Furthermore, for patients who receive oral mucosa graft Urethroplasty (OMGU), a smoking history will make the general state of oral mucosa worse, leading to poor graft survival after OMGU operation and ultimately leading to an increase in stricture recurrence rate. In urethroplasty using oral mucosa as a substitution, the effect of tobacco on the oral mucosa must also be considered. Although, in some studies, patients have been advised to avoid smoking before taking oral mucosal materials, long-term smoking history can hardly avoid the impact on the viability of oral mucosa, which may further increase the impact of smoking on oral mucosa and the recurrence of stricture.In some current urological guidelines, the effect of smoking on the stricture recurrence after urethroplasty is not mentioned. Although EAU related narratives had clearly mentioned that smoking had an important influence on the choice of surgery, in the follow-up part, it was still not mentioned that smoking is an important risk factor for postoperative recurrence.According to the results of this meta-analysis, urologists should guide urethroplasty patients to quit smoking before and after the operation to improve the overall success rate. Some potential limitations of this study should be presented. First, although some prospective data was involved, all the included studies are retrospective. Second, although it was recognized in statistical methodology, it is still possible to bring some additional bias by combining HR and OR to get RR estimates. Third, since smoking can directly damage oral mucosa, OMGU patients with a smoking history may have a higher recurrence ratio. It is also worth noting that the many other factors, such as postoperative complications and other factors, obviously play an important role in the recurrence of strictures, but not every primary study has fully adjusted the effects of these factors on smoking. However, in this meta-analysis, since many studies didn't offer detailed information about the OMGU technique, so OMGU subgroup analysis was not performed, further high-level evidence about smoking's effect on OMGU is needed.Our study shows that smoking can increase stricture recurrence risk after the urethroplasty. Quitting smoking may be a good option for patients undergoing urethroplasty surgery.


CNTNAP2, MTHFR, OXTR, SLC25A12, and VDR were identified, of which associations with suggestive evidence included the C677T polymorphism of MTHFR  and the rs731236 polymorphism of VDR . Associations with weak evidence included the rs2710102 polymorphism of CNTNAP2 , the rs7794745 polymorphism of CNTNAP2 (under dominant and heterozygote models), the C677T polymorphism of MTHFR (under homozygote model), and the rs731236 polymorphism of VDR (under dominant and recessive models). Our UR summarizes research evidence on the genetics of ASD and provides a broad and detailed overview of risk genes for ASD. The rs2710102 and rs7794745 polymorphisms of CNTNAP2, C677T polymorphism of MTHFR, and rs731236 polymorphism of VDR may confer ASD risks. This study will provide clinicians and healthcare decision-makers with evidence-based information about the most salient candidate genes relevant to ASD and recommendations for future treatment, prevention, and research.Autism spectrum disorder (ASD) is a class of neurodevelopmental conditions with a large epidemiological and societal impact worldwide. To date, numerous studies have investigated the associations between genetic variants and ASD risk. To provide a robust synthesis of published evidence of candidate gene studies for ASD, we performed an umbrella review (UR) of meta-analyses of genetic studies for ASD (PROSPERO registration number: CRD42021221868). We systematically searched eight English and Chinese databases from inception to March 31, 2022. Reviewing of eligibility, data extraction, and quality assessment were performed by two authors. In total, 28 of 5062 retrieved articles were analyzed, which investigated a combined 41 single nucleotide polymorphisms (SNPs) of nine candidate genes. Overall, 12 significant SNPs of  Autism spectrum disorder (ASD) is a group of neurodevelopmental conditions characterized by early-onset dysfunctions in communication, impairments in social interaction, and repetitive and stereotyped behaviors and interests . PatientDue to the lack of clinical and epidemiological evidence for an ASD cure, researchers have focused on better understanding ASD and advancing risk prediction and prevention . The cauTo date, numerous studies investigating the association between genetic variants and ASD risk have been published –11. MostAn UR collects and evaluates multiple systematic reviews and meta-analyses conducted on a specific research topic, provides a robust synthesis of published evidence, and considers the importance of effects found over time . In addiWe systematically searched the PubMed, EMBASE, PsycINFO, Web of Science, Cochrane Library, China National Knowledge Infrastructure, Sinomed, and Wanfang databases from inception to March 31, 2022. The databases were searched using the following strategy:  AND (gene* [All Fields] OR genom* [All Fields]) AND . Authors S. Qiu and Y. Qiu independently conducted literature searches for potential articles included in this review. The references of the relevant articles were manually searched to identify and incorporate eligible studies.We included meta-analyses of family based and case-control studies that examined associations between ASD and potential risk genes. We only included meta-analyses that reported either effect estimates of individual study or the data necessary to calculate these estimates. We excluded meta-analyses if (1) risk genes were used for screening, diagnostic, or prognostic purposes; (2) a study examined ASD as a risk factor for other medical conditions; (3) a study included fewer than three original studies investigating the association between risk genes and ASD; and (4) a study with missing information after the corresponding author, whom we contacted through email, failed to provide the required information. All articles retrieved were first organized in the reference manager software , and duplicates were deleted. S. Qiu and Y. Qiu chose eligible articles by screening the titles, abstracts, and full article texts independently. Disagreements were resolved through a discussion with a third investigator (Y. Li) until a consensus was reached.OR]) with the corresponding 95% confidence interval (CI), sample size of cases and controls, genotype and allele counts, and individual study designs . We used the ‘assessment of multiple systematic reviews’ tool, consisting of 11 items, to assess the methodological quality of the meta-analyses [From each eligible meta-analysis, we extracted the first author, publication year, genetic risk factors examined, number of studies, number of ASD cases and participants, study-specific relative risk estimates  to explain the between-study heterogeneity and to assess the uncertainty of a new study [I2-statistic [P > 0.10, I2 ≤ 50%), data were pooled using a fixed-effect model; otherwise, heterogeneity was evaluated using a random-effect model [P-values < 0.10 were considered to indicate the presence of small-study effects [P < 0.05) was greater than the expected number [P < 0.05, except where otherwise specified.In agreement with previous URs, we performed a statistical analysis using a series of tests that were previously developed and reproduced , 15, 16.CI . We alsoew study , 19. Hettatistic , 21. If ct model . The Har effects , 24. Thed number , 25. AllP-values by random effects model, small-study effects, excess significance bias, heterogeneity (I²), and 95% CI.In line with previous URs, we categorized the strength of the evidence of risk genes for ASD into five levels: convincing (class I), highly suggestive (class II), suggestive (class III), weak (class IV), and not significant , 26–28. This review was prospectively registered with PROSPERO (registration number: CRD42021221868).A total of 5062 articles were identified through an initial search. After removing duplicates, the titles and abstracts of 3182 articles were screened for eligibility. Of the remaining 66 articles that were reviewed in full, 28 eligible articles were selected for data extraction  [SLC6A4) [CNTNAP2) [OXTR) [RELN) [GABRB3) [SLC25A12) [VDR) [COMT) [COMT and SLC6A4). These studies were published from 2008 to 2021 and considered the associations between 41 SNPs in nine candidate genes and ASD risk. For quality assessment, 22 articles that scored 5−8 were rated as ‘moderate quality’, and six that scored < 5 were rated as ‘low quality’. Seventeen studies (60.7%) performed the HWE check (Table The characteristics of the selected studies are presented in Table MTHFR) –36; four[SLC6A4) –40 and cCNTNAP2) –44; thre) [OXTR) –47 and r) [RELN) –50; two [GABRB3) , 52, solLC25A12) , 54, and2) [VDR) , 56; and) [COMT)  , homozygote , and recessive  models. In addition, we found that the mutant allele of rs7794745 (CNTNAP2) increased ASD risk based on the dominant  and heterozygote  models. The C677T polymorphism of MTHFR was associated with an increased ASD risk in the allelic , dominant , heterozygote , and homozygote  models. The rs607755 polymorphism of RELN was associated with an increased ASD risk in the allelic , dominant , heterozygote , and homozygote  models. The rs731236 polymorphism of VDR was associated with an increased ASD risk in the allelic , dominant , homozygote , and recessive  models. In addition, we found that the mutant allele of rs7975232 (VDR) decreased ASD risk  based on the allelic model. There was no significant association between the other SNPs and ASD risk  under the allelic, homozygote, and recessive models; rs7794745 (CNTNAP2) under the heterozygote model; rs607755 (RELN) and rs731236 (VDR) under the allelic and homozygote models (Table MTHFR) under the allelic, dominant, and heterozygote models; a meta-analysis on C677T (MTHFR) under the five genetic models; a meta-analysis on rs20317 (GABRB3) under the dominant and heterozygote models; one each on rs736707 (RELN) and rs1544410 (VDR) under the recessive and allelic models, respectively; and three meta-analyses on rs607755 (RELN), 5-HTTLPR (SLC6A4), and rs7975232 (VDR) under the heterozygote model (P < 0.10).As for the results of CNTNAP2) under the allelic, homozygote, and recessive models; rs4680 (COMT) under the allelic model; rs20317 (GABRB3) under the heterozygote model; A1298C (MTHFR) under allelic, dominant, heterozygote, and recessive models; C677T (MTHFR) under homozygote and recessive models; rs736707 (RELN) under allelic, dominant, and homozygote models; 5-HTTLPR (SLC6A4) under allelic and recessive models; rs11568820 (VDR) under the dominant model; and rs731236 (VDR) under the heterozygote model, with statistically significant (P < 0.05) excess of positive studies , the P-value based on the random effects model was significant at P < 0.05 under allelic, homozygote, and recessive models. Between-study heterogeneity was not significant , the 95% PI did not exclude the null value, and there was no excess significance bias (P > 0.05) under the five genetic models. For rs7794745 (CNTNAP2), the P-value based on the random effects model was significant at P < 0.05 under dominant and heterozygote models. For C677T (MTHFR), there was a total of 2147 ASD cases, which was > 1000, and the P-value based on the random effects model was significant at P < 10–3 under allelic, dominant, and heterozygote models. Moreover, it was significant at P < 0.05 under the homozygote model. Between-study heterogeneity was large (I² > 50.0%) under the five genetic models, the 95% PI did not exclude the null value under the five genetic models, and there was no excess significance bias (P > 0.05) under allelic, dominant, and heterozygote models. For rs731236 (VDR), there was a total of 1088 ASD cases, which was >1000, the P-value based on the random effects model was significant at P < 10–3 under allelic and homozygote models, and the P-value was significant at P < 0.05 under dominant and recessive models. Between-study heterogeneity was not significant , the 95% PI excluded the null value, and there was no small-study effect (P > 0.10) and excess significance bias (P > 0.05) under the five genetic models (Table CNTNAP2) was graded as weak evidence (class IV) under allelic, homozygote, and recessive models; rs7794745 (CNTNAP2) was graded as weak evidence (class IV) under dominant and heterozygote models; the C677T (MTHFR) was graded as suggestive evidence (class III) under allelic, dominant, and heterozygote models; C677T (MTHFR) was graded as weak evidence (class IV) under the homozygote model; VDR (rs731236) was graded as suggestive evidence (class III) under allelic and homozygote models; and VDR (rs731236) was graded as weak evidence (class IV) under dominant and recessive models.We categorized the strength of the evidence of 20 SNPs for ASD into five levels. According to the criteria for the level of evidence, for rs2710102  were the C677T polymorphism of MTHFR  and rs731236 polymorphism of VDR . Associations with weak evidence (class IV) were the rs2710102 polymorphism of CNTNAP2 , rs7794745 polymorphism of CNTNAP2 (under dominant and heterozygote models), C677T polymorphism of MTHFR (under homozygote model), and rs731236 polymorphism of VDR (under dominant and recessive models).This UR summarizes evidence on the genetic basis of ASD. Our study design provides a robust and significant synthesis of published evidence and increases the conclusive power with more precise estimates. Overall, 12 significant SNPs of ASD remains a ‘disease of theories’, as multiple genes and environmental risk factors are probably involved in its pathogenesis. However, to date, the etiology and pathological mechanism of ASD are still unknown . The genCNTNAP2 is a member of neurexin superfamily and is a synaptic protein [CNTNAP2 mutations may be linked to the abnormal behavior of ASD by altering synaptic neurotransmission, functional connectivity, and neuronal network activity [CNTNAP2, with four and three meta-analyses reporting the associations with ASD, respectively. The results of the meta-analysis by Uddin et al. were inconsistent with the other authors’ [CNTNAP2 were associated with decreased risk of ASD. The rs2710102 was graded as having a weak association with ASD under allelic, homozygote, and recessive models. The rs7794745 was graded as having a weak association with ASD under dominant and heterozygote models. Therefore, it is likely that the rs2710102 and rs7794745 polymorphisms of CNTNAP2 influence the risk of ASD. protein . It play protein . CNTNAP2activity , 62. Theauthors’ . We furtMTHFR is one of the most frequently-researched genes in ASD, with four and eight meta-analyses for A1298C [MTHFR are associated with reduced enzymatic activity, which affects folate metabolism, and, consequently, fetal brain development [MTHFR has been the focal point of investigation in this disorder. The meta-analysis by Li et al. was selected because it was the most recent among the examined meta-analyses [MTHFR in the control group were not found in the HWE, which may be due to selection bias, population stratification, and genotyping errors within the original studies. We found no significant association between the A1298C polymorphism of MTHFR and ASD risk in the five genetic models, which was consistent with the four meta-analyses, indicating that the A1298C polymorphism of MTHFR may not be a risk SNP of ASD. We found that the C667T polymorphism of MTHFR was associated with an increased risk of ASD, graded as having suggestive association under allelic, dominant, and heterozygote models and weak association under the homozygote model. Thus, the C667T polymorphism of MTHFR may confer ASD risk.r A1298C –33 and Cr A1298C  polymorpelopment , 33. Dysanalyses . The genOXTR, a neuropeptide gene, is also one of the most frequently-studied genes associated with ASD [with ASD . Oxytociwith ASD , 63. Thrwith ASD , 46, andwith ASD , 46; howRELN encodes a large secreted extracellular matrix protein considered to be involved in neuronal migration, brain structure construction, synapse formation, and stability during neurodevelopment [RELN and ASD with conflicting results. None of the three meta-analyses found significant associations [RELN and ASD risk [P > 0.10, I2 ≤ 50%), a fixed model was applied to pool the effect size. We found that the rs607755 of RELN was associated with ASD risk in allelic, dominant, heterozygote, and homozygote models. This inconsistent result was caused by different pooling methods, indicating that it is necessary to perform an UR to provide a robust synthesis of published evidence and evaluate the importance of genetic factors related to ASD. Our UR results showed that the rs607755 of RELN was not significant when we categorized the strength of the evidence. Thus, it may not be a risk factor for ASD.elopment . Fatemi elopment . Dysfuncelopment , 65. Genciations –50. The SLC25A12 encodes the mitochondrial aspartate/glutamate carrier of the brain, a calcium-binding solute carrier located in the inner mitochondrial membrane that is expressed principally in the heart, brain, and skeletal muscle [SLC25A12. However, the results on the association between SNPs of SLC25A12 and ASD risk are inconsistent. Two meta-analyses were performed by Aoki et al. [l muscle , 67. Rosl muscle . Thus, ai et al.  and Liu i et al. , and desVDR gene, to which changes may be an underlying risk factor for ASD. Sun et al. [VDR and ASD risks. We further re-analyzed and categorized the strengths of evidence. The rs731236 polymorphism of VDR was associated with an increased risk of ASD, graded as having a suggestive association under allelic and homozygote models and a weak association under dominant and recessive models without small-study effects, excess significance bias, and large heterogeneity. It is likely that the VDR rs731236 polymorphism influences the risk of ASD.Vitamin D plays a significant role in brain homeostasis, neurodevelopment, and immunological modulation, and its deficiency has been reported in children with ASD . Hence, n et al.  and Yangn et al.  performeOur study has some limitations. First, associations between several SNPs and ASD risks under five genetic models or in different populations were not fully assessed in our UR, partly due to insufficient original data. Second, our UR is limited by significant heterogeneity that may be caused by population stratification, study design, and differences in the pattern of linkage disequilibrium structure. Finally, ASD is a complex disorder with different causative factors . We did not investigate the involvement of environmental factors in ASD. Despite these limitations above, our UR includes its prospective registration with PROSPERO, an extensive search strategy, clear criteria of inclusion and exclusion, duplicated processing by two authors, accurate quality assessment, systematic assessment and critical comparison of meta-analyses, and consistent standards for re-analysis of original data.CNTNAP2, C677T polymorphism of MTHFR, and rs731236 polymorphism of VDR may confer ASD risk. This study will aid clinicians in decision-making through the use of evidence-based information on the most salient candidate genes relevant to ASD and recommendations for future treatment, prevention, and research.In conclusion, our UR summarizes evidence on the genetics of ASD and provides a broad and detailed overview of risk genes for ASD. The rs2710102 and rs7794745 polymorphisms of


Retinol-binding protein 4 (RBP4) has been considered to be related to metabolic related diseases, such as hyperuricemia, obesity, and diabetes mellitus. However, whether nonalcoholic fatty liver disease (NAFLD) is related to RBP4 is unclear. Previous studies on the relationship between NAFLD and RBP4 levels have yielded inconsistent results. Hence, this meta-analysis was aimed to clarify whether circulating RBP4 levels are in relation to the risk of NAFLD.A meta-analysis was performed by applying observational studies to evaluate circulating RBP4 levels and NAFLD. Eligible studies published up to September 23, 2022, were searched in Embase, PubMed, and Cochrane databases.2: 89.8%). This association was confirmed in the Yellow race. However, no significant association was noted in the Caucasian race. After excluding the morbidly obese Population from the weight loss study (n = 2), the results of the comparison remained largely unchanged . Remarkable publication bias was not found. Although considerable heterogeneity was observed among the studies, no potential sources of heterogeneity were found in the subgroup analysis. Diagnostic methods for NAFLD were determined to be a potential source of statistical heterogeneity in meta-regression.In this study, 17 cross-sectional studies involving 8423 participants were included. Results from a random effects model showed that circulating RBP4 levels were higher in NAFLD patients than non-NAFLD (standardized mean difference (SMD) 0.28; 95% confidence intervals (CI): 0.11–0.46, IThe findings provide evidence that NAFLD patients exhibit higher levels of circulating RBP4 compared with controls, but high heterogeneity was observed. Thus, a high RBP4 level is probably a potential risk factor for NAFLD. To confirm the causal link between NAFLD and RBP4 level of causality, further prospective cohort studies are needed. Non-alcoholic fatty liver disease (NAFLD) is a major public health issue impacting about one quarter of the global population  and its Retinol binding protein 4 (RBP4) is a 21 kDa protein belonging to the Lipocalin family and is the only known particular transporter protein for vitamin A, which regulates the circulating levels of retinol . RBP4 isTill date, the etiology and pathogenesis of NAFLD has not been fully elucidated. Numerous studies have indicated that it involves an intricate interaction among obesity, environmental factors, microbiota changes, and susceptibility gene variants –15. ThesThis meta-analysis is registered on prospero (ID: CRD42022362281).Two independent reviewers (Rui Hu and Xiaoyue Yang) applied the subject terms “non-alcoholic fatty liver disease” and “retinol binding protein 4” and their corresponding free words in the Embase, PubMed, and Cochrane database for studies published before September 23, 2022. Without setting any language restrictions, all potentially eligible studies were included. In addition, the review articles and reference lists in the relevant original studies were examined to search for other potentially eligible researches. Based on the search strategy, all literature collected was independently assessed by two authors (Rui Hu and Xiaoyue Yang). They were both blinded to each other’s records and sought out potentially eligible studies. Any issues raised were addressed through discussion with the fourth author, Guangyao Song. A flow chart depicting the literature selection methodology is given in Fig. Inclusion criteria: (1) observational researches involving humans; (2) age ≥ 18 years; (3) studies involving measurement of the circulating RBP4 levels in NAFLD and non-NAFLD patients.Exclusion criteria: Past studies wherein the participants had liver injury due to infectious factors (such as hepatitis B or hepatitis C virus), total parenteral nutrition, alcohol use, drug use, or genetic causes. After excluding these articles, those with insufficient data or those studies that did not include non-NAFLD population were excluded.Two researchers (Rui Hu and Xiaoyue Yang) used a specially designed form extracted the independently the necessary data. Any discrepancies were resolved by discussion or by involving a fourth reviewer (Guangyao Song) whenever deemed necessary. The extracted data included the publication year, first author’s name, the design of study, participants’ characteristics , NAFLD diagnosis methods, and the technique of measuring RBP4. To sought the missing details, the appropriate person was contacted.Two researchers (Rui Hu and Xiaoyue Yang) assessed the extracted researches. The modified Newcastle–Ottawa Quality Assessment Scale (NOS)  was applP-value was extracted in this meta-analysis.Stata14.0 software was applied to conduct all statistical analyses. The standardized mean difference (SMD) was applied to evaluate the study outcomes. In addition, the mean ± standard deviation (SD) was expressed as the normally distributed data. Considering the potential for significant heterogeneity between the enrolled studies, this meta-analysis opted for a more conservative random-effects model. Two-tailed 2 statistics, the heterogeneity was evaluated. For Q statistics and I2 statistics, P < 0.10 and I2 < 50% was considered as statistical significance. An I2 values of 25% was poor heterogeneity, 50 and 75% represented medium and high heterogeneities, respectively [Using Q and Iectively .Through subgroup analyses, the influence of the potential factors on the correlation was assessed, which included ethnicity, the method of diagnosing NAFLD, and the method of measuring the RBP4 levels. In addition, possible sources of heterogeneity were analyzed using meta-regression.The stability of the outcomes was evaluated for sensitivity analyses by removing individual studies. In addition, Egger’s, Begg’s tests, and funnel plot were applied to assess the publication bias.The initial literature search resulted in the identification of 597 articles. After removing 87 duplicates and 467 irrelevant articles, 43 articles remained. After full text screening, 28 articles were excluded because of disagreement with the eligibility criteria listed in Fig. The features of the studies included in this article were shown in Table A modified NOS scale was used to assess the quality of the included studies. As presented in Table 2: 89.8%, P = 0.001) . However, circulating RBP4 levels in Caucasian NAFLD patients were not statistically different from those of healthy controls . Nevertheless, there remained a high level of heterogeneity in the comparison between Yellow  and Caucasian  populations.To examine whether differences in ethnicities  could be the source of the high subgroup analysis was conducted according to the different ethnicities included in the study. As shown in Fig. P = 0.000). However, in the subgroup with liver biopsy as a diagnostic method, patients with NAFLD had no statistically significant difference in circulating RBP4 levels compared to non-NAFLD . These results suggest that the circulating RBP4 was higher in patients with abdominal ultrasound as a diagnostic method than in controls.  , only a small change was noted in the comparison between NAFLD patients and without NAFLD   Fig. . FurtherP = 0.002; multivariate meta-regression: P = 0.024).In the meta-regression analysis Table , the diaP = 0.217)  Figs. , 9, 10.FP > 0.05) disagree with the results of this one [In this meta-analysis, higher levels of circulating RBP4 were found in NAFLD patients compared to non-NAFLD, although heterogeneity between the combined studies was significant. The findings of a previous meta-analysis (SMD = 0.08, 95% CI: − 0.21 to 0.38, this one . Howeverthis one . In contthis one . Thus, tThe mechanism by which RBP4 causes NAFLD progression remains unclear. Liu  found viIn this meta-analysis, high heterogeneity between studies was found when performing subgroup analyses. Moreover, meta-regression suggested that the diagnostic method of NAFLD may be a potential source of heterogeneity. After subgroup analysis based on the diagnostic method of NAFLD, the heterogeneity reduced slightly. Nonetheless, there were no remarkable differences in NOS scores, patient age, sex, measurement method, or diagnostic methods for NAFLD before and after the exclusion of outliers; hence, these results are robust. During the measurement of RBP4 level, only some studies grouped the values according to the severity of NAFLD, Zhou  did not PNPLA3 and the 167Lys allele of TM6SF2 have been shown to be strongly linked to the increase in hepatic fat deposition and the development of cirrhosis and hepatocellular carcinoma [Genetic variants in the I148Met allele of arcinoma . Additioarcinoma , 46 havearcinoma . Only a arcinoma  showed tThe current meta-analysis including 17 clinical researches has a larger sample size than previous meta-analyses and includes literature from a wider region, including Asia, Europe, and Africa , 37, 38.However, some limitations are present. First, all the included studies were cross-sectional studies, and the causal relationship between RBP4 levels and NAFLD could not be established. Second, although the indicators of circulating RBP4 levels were converted to the same units, the values varied from one study to the other. For patients with NAFLD, the highest RBP4 level was 94.6 μg/mL , whereasPatients with NAFLD exhibited higher levels of RBP4 than healthy controls. Hence, RBP4 may be an emerging noninvasive biomarker for NAFLD. In addition, RBP4-targeting agents can be considered for the treatment of this disease.


Several dietary exposures have been associated with gastric cancer (GC), but the associations are often heterogenous and may be afflicted by inherent biases. In the context of an Umbrella Review (UR), we provide an overview and a critical evaluation of the strength and quality, and evidence classification of the associations of diet-related exposures in relation to the risk of GC. We searched PubMed and Scopus for eligible meta-analyses of observational studies published in English from inception to 12 December 2021, and for any identified association, we applied robust epidemiological validity evaluation criteria and individual study quality assessment using AMSTAR. We screened 3846 titles/abstracts and assessed 501 full articles for eligibility, of which 49 were included in the analysis, investigating 147 unique exposures in relation to GC, cardia (GCC) or non-cardia (GNCC) cancer. Supported by suggestive evidence, positive associations were found comparing the highest vs. lowest categories for: heavy (>42 g/day) alcohol consumption (Relative Risk (RR) = 1.42, 95% Confidence Interval (CI): 1.20–1.67), salted fish consumption  and waist circumference  and an inverse association for the healthy lifestyle index  in relation to GC. Additionally, a positive association was found comparing obese individuals (Body Mass Index (BMI) ≥ 30) to normal-weight individuals (BMI: 18.5–25)  in relation to GCC. Most of the meta-analyses were of medium-to-high quality . Maintaining a normal body weight and adopting healthy dietary choices, in particular, limiting the consumption of salt-preserved foods and alcohol, can reduce the risk of gastric cancer. Gastric cancer (GC), also known as stomach cancer, is among the five most common cancers worldwide, accounting for more than 750,000 deaths in 2020 . Most ofThe global rates of gastric non-cardia cancer (GNCC) are declining, which can be attributed to a lower prevalence of Helicobacter pylori infection (H. pylori), a well-established risk factor for GNCC . On the The aim of the present study is to provide an overview and a critical evaluation of the strength and quality, and classification of the existing epidemiological evidence that investigates the association of diet-related exposures in relation to risk of GC, through the prism of an umbrella review, and, where possible, to highlight methodological gaps and reasons for inconsistency.https://osf.io/pd5br/).This umbrella review follows the proposed reporting standards, using publicly available checklists ,10. FurtTwo reviewers independently searched MEDLINE via PubMed and Elsevier’s Scopus for eligible meta-analyses published in English from inception to 12 December 2021 using diet- and gastric-cancer-related terms .Eligible for inclusion were meta-analyses that examined the association between any diet-related exposure in relation to GC risk. No restrictions were made regarding population characteristics, such as age, gender, ethnicity or setting. Studies that only examined survival or recurrence, or GC mortality (with no consideration of first onset GC risk endpoints), and narrative reviews were excluded from the review.Title, abstract and full text screening was performed independently by two reviewers (EB and AS), using a publicly available systematic review tool (Rayyan), and potential discrepancies were resolved by a third reviewer (ABH) . In the Data were extracted by two reviewers (AS and MT) using a predefined data extraction form and verified by another reviewer (EB). From each meta-analysis, the following items were extracted: definition of exposure and type of contrast , sample size (number of cases and controls), effect size of the association , along with 95% confidence intervals (CI)) and design of each primary study.Where available, analyses were performed separately using primary studies with a prospective design and case–control studies. In the case that both the prospective and case–control studies were available for a given association, the prospective ones were included in the main analysis, and the case–control one were analyzed separately in a sensitivity analysis.2) was used, and the prediction intervals (PI) were estimated to further quantify the heterogeneity [The effect size of each association was estimated using a random effects meta-analysis, by means of RR along with 95% CI . The incogeneity ,14. Smalogeneity ,16. SummThe description of the included studies in the present umbrella review was also supported using describing bibliographic information as provided by Scopus, such as paper author, journals, affiliated countries and citations .All statistical analyses were performed using the libraries meta, metafor & bibliometrix in R v. 4.1.0. ,18,19,20p-value < 0.05) was classified as strong evidence (Class I), highly suggestive evidence (Class II), suggestive evidence (Class III) or weak associations (Class IV), with regard to epidemiological validity criteria ; 95% prediction intervals excluding the null; and no evidence of small study effects and excess significance bias. Furthermore, the quality of the included studies was evaluated using AMSTAR [Each unique association that was nominally significant . Separate analyses were performed including both prospective and non-prospective (case–control) studies.Biases, such as selection and information bias, are commonly distorting the associations in observational studies, resulting in larger estimates with small The literature search yielded 3846 unique articles, of which 3345 were excluded during the title and abstract screening and a further 318 were excluded during the full-text screening. Of the 183 eligible studies, 49 systematic reviews with MA examining 147 unique exposures in relation to GC, GCC or GNCC were finally included in the umbrella review . The 49 The majority of the exposures , for which prospective primary data were available, were evaluated in relation to GC in the classification process; 15 were evaluated in relation to GCC and 14 in relation to GNCC . Sensitip-value < 0.05): 2 (1.7%) had p-values less than 10−6; 86 (74.8%) were based on more than 1000 cases; 85 (73.9%) had low heterogeneity (I2 < 50%); 3 (2.6%) had significant prediction intervals; 83 (72.2%) did not have evidence of small study effects; and 88 (76.5%) had no evidence of excess significance bias, yielding no strong or highly suggestive associations (Classes I and II), but 4 were suggestive evidence (Class III) (The 115 associations with GC were based on a median of 3 primary studies (range: 2–24), 1586  cases and 642,876  total participants. Under a random effects MA, 24 of 115 associations (20.9%) were nominally significant (ass III) . In partass III) . A minimp-value less than 10–6; only 2 were based on more than 1000 cases; most (10/15) had low heterogeneity, but no association had significant prediction intervals; less than half (6/15) had no evidence of small study effects and almost all (14/15) had no evidence of excess significance bias. Only one association, comparing obese individuals (with BMI ≥ 30 kg/m2) to normal-weight individuals (BMI: 18.5–25), was supported with suggestive evidence  (The 15 associations with GCC were based on a median of 2 primary studies (range: 2–7), 547 (96–3666) cases and 203,437  participants. Under a random effects MA, less than a third (4/15) of the associations were nominally significant, while no association had 32–2.49) , for whiThe 14 associations with GNCC were based on a median of 3 primary studies (range: 2–8), 925 (192–8391) cases and 438,515  participants. Only one nominal association was found, but it was based on just two primary studies, hence providing little evidence for an association .p-value < 0.05) for GC, GCC and GNCC, respectively, under a random effects MA , alcohol, nitrites and red meat  and dairy for which a nominal inverse association (Class IV) was found based on prospective studies, but was non-significant in the sensitivity analysis .Quality assessment using AMSTAR showed that most of the studies were of medium to high quality . All the AMSTAR items were fulfilled in at least 70% of the included studies except for item 1 , item 2  and item 4  .Our analysis confirms a positive association for highly salted fish consumption, heavy alcohol drinking (>42 g/day) and abdominal obesity and an inverse association for a healthy lifestyle index, which includes healthy dietary choices in combination with other lifestyle factors  in relation to the risk of GC.High salt consumption, and particularly the high consumption of salt-preserved foods, such as fish, vegetables and meats, emerged as a probable GC risk factor in the 1970s . Ever si2 was found in a dose–response MA of seven prospective studies and a similar association was identified in the WCRF 2018 report, with no evidence of association for GNCC [Increased body fatness has been associated with 13 distinct types of cancer, including GCC, demonstrating the fact that it has both systemic and tissue-specific, cancer promoting effects . Obesityfor GNCC . In lineA 42% increased GC risk, supported by suggestive evidence, was found in our analysis for alcohol consumption at 42 g/day or higher, but no association was found, based on prospective evidence, for moderate habitual consumption in comparison to non-drinking. In a similar fashion, a significant linear dose–response association has been previously reported, only at levels of 45 g of ethanol per day and above . The obsPrevious studies have shown that the high consumption of fruit is associated with a decreased GC risk. Specifically, a 5% decreased GC and a 24% decreased GCC risk, per 100 g of fruit and citrus fruit consumption per day, respectively, have been previously reported ,35. The A combination of healthy lifestyle factors, captured by pertinent scales, was associated with a suggestive inverse association with GC. Diverse scoring systems were utilized in the primary studies, which included heterogenous populations, with different socio-economic backgrounds ,42,43. TWe provide a comprehensive analysis of a wide range of dietary exposures, using a rigorous evidence credibility evaluation methodology. Furthermore, by focusing on prospectively collected data from well-established cohort studies as our primary unit of analysis, we limit the extent to which the observed associations may be afflicted by biases, such as selection and information bias. Additionally, the large sample size that most of the studies were based on provides enough power to identify emerging associations. On the other hand, our study has several limitations, most of which stem from the primary studies: it lacks an evaluation of the linear or dose–response associations, and an evaluation of the associations by sex. Furthermore, dietary assessment using non-validated approaches, inadequate adjustments in some of the primary studies, residual confounding and the absence of follow-up dietary assessments to capture potential changes in the participants’ habits are some of the inherent limitations. It should be noted that H. pylori infection was not considered in the vast majority of the primary studies. A widespread testing for and eradication of H. pylori infection is not supported; however, gaining a better understanding of the determinants of host susceptibility to GC, post infection, is important to inform prevention strategies. Estimates across meta-analysis were not pooled in the present umbrella review; hence, primary study overlapping is not considered a source of bias. Nevertheless, participant overlapping  may have occurred to an extent, inducing a correlation of the MA estimates, but this is unlikely to have biased the associations substantially.In conclusion, maintaining a normal body weight and adopting healthy dietary choices, in particular, limiting consumption of salt-preserved foods and alcohol, in combination with other lifestyle factors, such as smoking abstinence and regular physical activity, reduce the risk of developing gastric cancer.


Toll-like receptor 4 (TLR4) is a lipopolysaccharide receptor that may influence tumor progression through inflammatory response and immune response. This complex process mainly occurs within cells. The correlation between TLR4 and neoplasms has been of great interest, but discrepancies remain.We analyze the literature retrieved from five databases  to assess the intensity of association using odds ratio (ORs) and 95% confidence intervals (95% CI). Meta-regression and subgroup analysis were utilized to find sources of heterogeneity. Publication bias is estimated using contour-enhanced funnel plots, Begg’s test, and Egger’s test, and we implemented sensitivity analysis to clarify the reliability of the outcomes. We also conducted an evaluation of the sample size using trial sequential analysis (TSA) method.We found a significant association between rs4986790 and tumors (dominant model: OR [95% CI] = 1.25 [1.11–1.42]; heterozygous model OR [95% CI] = 1.25 [1.11–1.41]; and additive model: OR [95% CI] = 1.25 [1.10–1.41]. Specifically, the rs4986790 minor allele G may increase the risk of gastric cancer (dominant model: OR [95% CI] = 1.62 [1.3–2.03]; heterozygous model: OR [95% CI] = 1.57 [1.24–1.97]; additive model: OR [95% CI] = 1.64 [1.31–2.05] and other tumors (dominant model: OR [95% CI] = 1.36 [1.17–1.57]; heterozygous model: OR [95% CI] = 1.43 [1.25–1.63]; additive model: OR [95% CI] = 1.35 [1.18–1.55]. Further subgroup analysis showed that this association are both present in Caucasian and Asian.The outcomes of our systemic review proved that the TLR4 polymorphism rs4986790 is associated with cancer, especially with gastric cancer, and this strong correlation are evident in Caucasians and Asian. In 2020 alone, about 20 million new cancer patients and about 10 million cancer deaths are expected worldwide. Although cancer therapy is now more efficient globally, the high incidence and mortality of cancer cannot be ignored. The genetic risk of most cancers is due to multiple risk alleles, each with low to moderate risk. Although rare, mutations in the single nucleotide polymorphisms minor allele are likely to be linked to the development and progression of cancers. Tumor progression involves complicated interplays between tumor cells, immune cells, and the tumor microenvironment. Studies have shown that persistent infection and chronic inflammation can promote many malignancies, such as colorectal, gastric, breast, uterine, liver, and prostate cancers (PCs).–11 The potential connection opens up new possibilities for treating cancers associated with chronic persistent infections.Cancer has become a common global burden to be alleviated due to its high mortality rate, and cancer has become a severe threat to human health.,13 All TLRs are architected as Toll receptors consisting of an external pathogen-associated molecular patterns recognition domain with leucine-rich repeat motifs and an intercellular signaling domain.,15 The signaling pathway of TLRs in human individuals originates from the TIR structural domain of the cytoplasm, which is conserved in all TLRs. Due to the resemblance of the cytoplasmic portion of TLRs to the interleukin-1 receptor family, TIR is also known as the Toll/interleukin-1 receptor structural domain. Studies have shown that adapters such as TIRAP, Myd88, TRIF, and TIRP contain TIR domains that modulate the TLR signaling. Myd88 may affect lipopolysaccharide (LPS)-triggered inflammatory responses through adverse regulatory effects. TIRAP binds to TLR4 and participates in the TLR4-mediated Myd88-independent signaling. Compared to MyD88 or TIRAP, overexpression of TRIF triggers IFN-promoter activation and participates in the TLR3-mediated MyD88-independent pathway. TIRP, on the other hand, is a recently discovered adapter containing a TIR domain.Toll-like receptor 4 (TLRs) are family members of known transmembrane glycoproteins, 13 identified in humans. As a vital member of the TLRs family, TLR4 has an extracellular structural domain composed of 22 leucine-rich repeat sequences, which can be expressed in some cells. Upon activation by LPS, LBP translocates to and binds to the TLR4/MD-2 complex via CD14. TIR domain-containing adapters are activated to translocate endotoxin into cells, promoting inflammatory factor production, and exerting its influence on tumorigenesis and development. The polymorphism rs4986790 substitutes the A allele at 896 base pairs by G, resulting in the substitution of glycine for aspartate at amino acid sequence (TLR4_896A/G) site 299. This gene missense mutation alters the extracellular configuration of this receptor and is linked to a blunted response to LPS in vitro and in vivo. Many current studies have suggested a potential relationship between TLR4 polymorphisms and tumors. Zhou et al, Zhao et al, and Zhang et al substantiated that TLR4 may increase gastric cancer (GC) susceptibility, and Li et al revealed that TLR4 plays a crucial role in colorectal carcinogenesis. However, some studies did not suggest an apparent association.–29 Because of these conflicting conclusions, we are undertaking the most extensive systematic review, enhancing study precision with large sample sizes.Drosophila Toll protein (TLR4) is a representative receptor for Gram-negative LPS of Gram-negative bacteria, which is a critical bridge molecule linking oncogenic infections to cancer.This meta-analysis was prospectively registered on PROSPERO. We developed the study protocol according to the PRISMA guidelines. This study does not directly involve animal or human trials and does not require an ethics approval statement.http://links.lww.com/MD/H680, Supplemental Digital Content, which demonstrates specific search strategies). We also ensured that some studies that might have been missed were selected by manually searching the reference citations included in the article. If the criteria are met, some of the gray studies will be included as study data, Including dissertation, preprint, Etc. If there is any inconsistency between two authors during the search process, a third author must be involved until the problem is resolved.Two authors searched the literature from five web databases , independently retrieved and selected appropriate records, and used Endnote software to categorize and organize related literature. The Searched pieces of literature are published until March 2022, without language or geographical restrictions. The search formula will comprise the below terms: “Toll-Like Receptor 4,” “TLR4 receptor,” “rs4986790,” “Neoplasia,” “Tumor,” “Cancer,” “Malignancy,” “Carcinoma,” “SNPs,” “Polymorphism,” and “Single Nucleotide Polymorphisms” and cancers risk; each tumor diagnosis meets the diagnostic criteria; had available genotype frequencies both in cases and controls. Articles that do not meet the appeal criteria will be excluded. Of the multiple studies for which duplicate data were published, only the most current or intact study was selected.P-values based on the χ2 test will be applied to weigh the HWE results, and studies with P > .05 will be consistent with HWE. Studies that do not comply with HWE will be excluded.Two investigators (QX and JC) involved in this study respectively extracted the following required contents from each of the 39 initially included papers : first author, publication year, country region, ethnicity, source of the control group, cancer type, and genotype frequency of case and control groups, genotyping method. The Hardy-Weinberg equilibrium (HWE) was calculated with data from each study control group. http://links.lww.com/MD/H681, Supplemental Digital Content, which aggregates the quality of all included literature).The quality results of the incorporated articles will undoubtedly affect the conclusions of our study. Therefore, we applied the Newcastle-Ottawa Scale (NOS) to estimate study quality. Two authors evaluated the incorporated studies separately, integrated the final results, and settled them by negotiation when inconsistencies arose; if essential, consider involving a third party until the issue was resolved. The evaluation form, with a total score of 9, is developed in three aspects: selection, comparability, and exposure. A total NOS score greater than six will be considered a high-quality study. Otherwise, it is of low quality  and 95% confidence intervals (CIs) to evaluate the connection between TLR4 polymorphism (rs4986790) and cancer riskQ-test and I-square to assess the heterogeneity level. P-values less than .1 or I-square greater than 50% were identified as high heterogeneity. For studies with high heterogeneity, we will apply a random-effect model to estimate the association between rs4986790 and cancers; otherwise, a fixed-effect model will be adopted. If heterogeneity exists, we will conduct meta-regression to find the origin of heterogeneity and further explore the specific origin of heterogeneity through subgroup analysis.We used a chi-square-based P-value of the Begg’s and Egger’s tests were less than 0.05.We apply sensitivity analysis to judge the reliability of the stability of the results. We eliminated sequentially one case-control study and analyzed the rest of the studies to compare whether the result varied from the initial result. If the results changed, then this study may have been excluded. This approach will make our study more credible. We evaluated the risk of publication bias with contour-enhanced funnel plots, Begg’s, and Egger’s tests. Publication bias was present if the white area of the funnel plot was asymmetric or if the P-value, statistical efficacy, and applicable prior probability. We set the FPRP threshold to 0.2 and examine the dominance OR of 0.67/1.50 (protection/risk effect) with a priori probability of 0.1. Results will be significant if the FPRP value is less than 0.2. The trial sequential analysis (TSA) results represent the sample size required for the current conclusions drawn from the meta-analysis. They can rectify the enhanced risk of type I and type II errors from multiple updates of the meta-analysis, which are mainly due to data merging. The application of TSA minimizes false-positive results resulting from random errors when the sample size of meta-analysis does not reach a sufficient sample size. The TSA software (0.9.5.10 Beta) is what we used. We run the software in strict accordance with the TSA user manual, which is available at www.ctu.dk/tsa. We used a type I error of 5%, a type II error of 20% (power of 80%), and the traditional cutoff value  of 1.96 (α = 0.05). The results were described synthetically by the relationship between the cumulative Z-value and the TSA boundary curve. If the cumulative Z-value reached the information size or transcended the TSA boundary curve, the sample size met the required number for the study; otherwise, the study required a more extensive original study to substantiate the results further.,34We evaluated the results using false-positive reporting probability (FPRP) analysis, where the FPRP value is determined by the observed http://gepia.cancer-pku.cn/) to analyze the differences in RNA sequencing expression data between tumor and normal tissues.–36 GEPIA data are derived from the TCGA and GTEx databases and can further explore the association between TLR4 and cancer. We present some of the results in the form of boxplots.We applied an open interactive website called GEPIA . We eliminated 83 duplicate records by literature management software (Endnote X9). Subsequently, we read the titles and abstracts of the literature and realized that there were still 49 articles that did not conform to the requirements. Full-text browsing of the remaining literature was continued, removing some of the literature, most of which had no acquired genotype frequencies. We also manually searched for reference citations in incorporated articles and similar articles. A summary of 19 articles was additionally selected. Researchers excluded three ineligible studies after assessing HWE based on control group data. The final number of included studies are 38  and Asian(n = 5). Some studies did not make a precise distinction between the ethnic groups, and we defined these as mixed ethnic groups (n = 2). The control population was obtained from the community and hospital by a randomized method. For the accuracy of the experiment, we have excluded studies that did not comply with HWE. In contrast, all included literature had quality assessment (NOS) scores above five, and all studies could be considered high quality , while no distinct heterogeneity was found in other subgroups , In contrast, this association is not distinctive in the mixed group population. And in the subgroup analysis of tumors, we found that rs4986790 is significantly associated with the risk of GC  and other cancers . For subgroup analyses of control group sources, we can conclude that population-based , hospital-based , and unknown populations  all manifested the connection between the TLR4 polymorphism rs4986790 and cancers.http://links.lww.com/MD/H682, Supplemental Digital Content, which demonstrates the results of quantitative analysis of publication bias in the included literature). Conducting sensitivity analyses can estimate the stability and credibility of study results. In the study, we found that the results (pooled ORs) changed after removing one study by Neamatallah et al in the recessive and homozygous models, so we excluded this study , and Z-curve intersects with the TSA boundary curve. However, it is worth mentioning that the number of cases corresponding to the cumulative Z value did not exceed the required information. In the sample size assessment of GC , and Z-curve intersects with the TSA boundary curve. In addition, the number of cases corresponding to the cumulative Z value exceeded the required information size.Table 6 α > 0.0, and Z-cThe results of in-silico analysis suggested that the expression of TLR4 in tumor tissues was significantly higher in acute myeloid leukemia than in normal tissues. Although the expression of TLR4 was more elevated in stomach adenocarcinoma tissues than in normal tissues, this discrepancy was not significant Fig. . Zhao et al, and Zhang et al.Our meta-analysis discussed the connection between TLR4 polymorphism (rs4986790) and tumors. We showed that rs4986790 increased the risk of GC, while no significant association was seen for colorectal cancer, PC, and non-Hodgkin’s lymphoma. This result is congruent with the conclusion of Zhou et al, It is worth affirming that the inflammatory response is an influential factor in cancer progression. Previously, TLR4 polymorphisms have been extensively studied, with rs4986790 being the most studied. Several studies have suggested an association between TLR4 and cancers, but some studies have not suggested the same results. Because of the inconsistency of these studies, we undertook this updated systematic review. This systematic review included 38 studies , the largest sample size to date, and the results are likely to be more credible. In addition, there were 13 tumor types included in our meta-analysis. Five studies examined the association of TLR4 polymorphism rs4986790 with PC, and six studies for non-Hodgkin’s lymphoma. There were 6 and 11 studies involving colorectal and GCs, separately. Others include acute myeloid leukemia, bladder cancer, breast cancer, cervical cancer, endometrial cancer, malignant melanoma, lung cancer, liver cancer, and chronic myeloproliferative neoplasm also included in our meta-analysis. However, each of them has less than two research articles. Explore as much as possible the relationship between tumors and TLR4.TLR4, as an LPS receptor, may promote inflammatory and immune processes through antigen-presenting cell responses.http://links.lww.com/MD/H683, Supplemental Digital Content, which shows a preliminary appraisal of sources of heterogeneity using meta-regression). Further subgroup analysis suggested significant heterogeneity in the Taqman subgroup, which may be due to potential genotyping method errors, the undeniable effects of chance errors, and experimental design methods; this may not be completely avoidable. The asymmetry in the white areas was found when assessing publication bias based on funnel plots. While neither Begg’s test nor Egger’s test showed publication bias, it led us to consider that this asymmetry was due to heterogeneity. In the meantime, we removed language restrictions during the literature record screening to minimize selection bias. For sensitivity analysis, we found an individual study in the Homozygous and recessive models that may have influenced the final results, which we excluded. Notably, due to the missing GG genotype of the population in some studies, ORs and 95% CIs could not be calculated, which may make the final combined ORs and 95% CIs less plausible.According to the previous results, the heterogeneity detected by meta-regression analysis mainly stemmed from the genotyping method  is more susceptible when the gene frequencies are low in the population. Wetterslev et al first introduced the concept of TSA and applied it in an updated meta-analysis, which then began to be widely applied.,77,78 Consequently, we examine the effect of random error on the results by TSA. The sample size was estimated to overcome the deficiencies of the study and minimize false-positive results. In the sample size assessment of all tumor types , reducing the number of studies included in both models that could be used for data analysis. The reason may be the low gene frequency of the minor allele G in the natural population (MAF: G = 0.0579). Finally, the presence of non-neoplastic diseases in a few study controls may impact the final results, specifically when these diseases correlate with the TLR4 polymorphism rs4986790.A broader range of races could be included in future studies while controlling for risk factors such as age or gender. Since potential biases and confounders cannot be wholly excluded throughout the study, we need larger, better-designed, and more comprehensive studies in the future. Research on TLR4 has been relatively advanced, including the association of TLR4 with cancer and the molecular mechanism of the TLR4 gene causing cancer. However, there are fewer studies on clinical efficacy, and further research is needed to determine whether the TLR4 gene can be a breakthrough in cancer treatment in the future. At the same time, we cannot overlook the role of TLR4 in tumor proliferation. Its mechanism of action in the inflammatory process of tumor development needs further to reveal, which can better serve the future of cancer prevention and therapy.There was a meaningful correlation between rs4986790 (A > G) and tumors. Among them, rs4986790 minor allele G may increase the risk of GC development. Further subgroup analysis revealed that this association exists in Caucasians and Asians.Thank you to Jiangxi Provincial Health Commission for the financial support.Conceptualization: Guomin Zhu.Data curation: Qiang Xiao.Formal analysis: Qiang Xiao, Jian Chen, Hu Cai.Funding acquisition: Guomin Zhu.Investigation: Jian Chen, Hu Cai.Methodology: Qiang Xiao.Project administration: Guomin Zhu.Resources: Jian Chen, ShuKun Zeng.Software: Qiang Xiao.Supervision: ShuKun Zeng.Validation: Jian Chen, ShuKun Zeng.Writing – original draft: Qiang Xiao.Writing – review & editing: Jian Chen.