Studies evaluating the association between monocyte chemoattractant protein-1 (MCP-1) -2518 A > G (rs1024611) polymorphism and type 2 diabetes mellitus (T2DM) and diabetic nephropathy (DN) are contradictory. The present study aims to provide a comprehensive assessment and more reliable estimation of the relationship between the MCP-1 rs1024611 polymorphism and T2DM and DN risk.Eligible articles were retrieved from the PubMed, Web of Science, EMBASE, Cochrane, and China National Knowledge Infrastructure databases. The effect summary odds ratios (ORs) and 95% confidence intervals (CIs) were obtained to calculate the summary effect size. Heterogeneity was analyzed by subgroup analysis and meta-regression. Publication bias was tested using funnel plots and Egger’s test.OR = 0.79, 95% CI: 0.66–0.93, P = 0.006; PQ = 0.372). No significant risks were found in the Asian population for any genetic models. Seven studies found an association between the GG + GA genotype and DN risk in the Asian population . No significant risks were found in the Caucasian population with any genetic models. There were no statistically significant differences in genotype distribution between patients with T2DM and DN in Asians or Caucasians. Meta-regression revealed that genotyping method was a major driver of heterogeneity in five genetic models .In total, sixteen studies were included. Thirteen studies involving 2,363 patients with T2DM and 4,650 healthy controls found no significant association between the MCP-1 rs1024611 polymorphism and T2DM in the overall population. Ethnicity stratification found an association between the GG + GA genotype and decreased T2DM risk in Caucasians (The MCP-1 rs1024611 polymorphism is associated with susceptibility to T2DM in Caucasians and DN in Asians. Larger, well-designed cohort studies are needed in the future to verify this association.The online version contains supplementary material available at 10.1186/s12902-023-01514-z. Type 2 diabetes mellitus (T2DM) is a major global health problem and its global prevalence is increasing annually . ApproxiThe etiology and pathogenesis of T2DM and DN have not been fully elucidated. Recently, the theory of inflammatory injury has received much attention. Evidence from many clinical and experimental studies shows that T2DM and DN are natural, immune, chronic, low-grade, inflammatory diseases –6. MonocIn recent decades, some case‒control studies have investigated the relationship between the MCP-1 rs1024611 polymorphism and the risk of DN or diabetes; however, the results have been inconsistent –29. StudThis meta-analysis followed the PRISMA guidelines for systematic reviews . SearcheStudies meeting all the following criteria were included: (1) observational studies ; (2) at least two comparison groups ; (3) the distribution of the genotypes in control group and T2DM group (when we compared T2DM group with DN group) was in Hardy–Weinberg equilibrium (HWE); and (4) providing the genotype distribution frequency of the DN and control groups , or the possibility of calculating this using the literature. Exclusion criteria included: (1) duplicate publications and (2) reviews, case reports, and meta-analyses.The PECO format was used , as follData extraction was performed independently by two researchers based on the inclusion and exclusion criteria. If any information was missing from the selected articles, we contacted the corresponding authors. The extracted information included the first author, year of publication, study design, country and ethnicity of subjects, HWE in the control group, and the number of genotypes in the case and control groups. The studies included in the analysis were scored according to the standard Newcastle–Ottawa Scale (NOS) . StudiesP < 0.05 was considered statistically significant. Subgroup analyses and meta-regression for the study population, genotyping method, comorbid chronic disease, and age- and sex-adjustment were performed to identify possible sources of heterogeneity.Stata version 12.0 software was used for all statistical analyses. A chi-square test was used to examine whether the distribution of genotypes among the control and T2DM groups (compared to the T2DM and DN groups) within each study was in HWE. The odds ratios (ORs) and 95% confidence intervals (CIs) were used to calculate the summary effect size and assess the association between the MCP-1 rs1024611 polymorphism and T2DM or DN susceptibility. Pooled ORs were calculated using the Z test for the dominant (GG + GA vs. AA), recessive (GG vs. GA + AA ), allele contrast (G vs. A), homozygote (GG vs. AA), and heterozygote models (GG vs. GA). I2 tests were used to evaluate statistical heterogeneity [PQ < 0.10 or I2 > 50%) was used to calculate the OR and 95% CIs; otherwise, the fixed-effects (FE) model was used to calculate combined effect estimates [Cochran’s (Q) and ogeneity . The ranstimates . A sensiA total of 462 studies were obtained through systematic searches. After reading the title, abstract and full text, 446 articles were excluded and the remaining 16 articles were included in the meta-analysis –29. A flThirteen studies with 2,363 patients with T2DM and 4,650 healthy controls were eligible and used to estimate the relationship between the MCP-1 rs1024611 polymorphism and T2DM. Significant heterogeneity was detected among the overall population in five genetic models  and RE model , without significant heterogeneity  , but no significant risk was found under the RE model , with significant heterogeneity . Significant associations with T2DM were found for the heterozygote model in Asians under the FE model , but no significant risk was found under the RE model , with heterogeneity   in studies without sex- or age-adjustment compared to studies with adjustment  is presented in Fig. OR = 1.27, 95% CI: 1.07–1.51, P = 0.007) under the FE model  ,without significant heterogeneity . No significant risks were found in the Caucasian population for any genetic model under either the FE or RE models   and in patients with DN with no comorbid chronic diseases    with 1,074 patients with DN and 1,361 patients with T2DM. Significant heterogeneity was detected in the overall and Asian populations under five genetic models , and there was a significant reduction in heterogeneity in the subgroup analysis for genotyping method.Meta-regression revealed that genotyping method was a major driver of heterogeneity in five genetic models . A funnel plot showed no significant asymmetry under the dominant model (GG + AG vs. AA) Fig. . P valueIdentification of the possible genetic origin of T2DM and DN could provide a theoretical basis for the early diagnosis or intervention of T2DM and DN. In this pooled analysis, we found that the GG + GA genotype was associated with a lower risk of T2DM in Caucasians, and that the GG + GA genotype was associated with DN risk in Asians. Subgroup analyses were conducted according to population, genotyping method, comorbid chronic disease, and age- and sex-adjustment. We conducted a detailed and comprehensive analysis under five genetic models and three comparative groups. No evidence of publication bias was observed under all genetic models. In addition, we adopted a sensitivity analysis, which did not affect the results of nonsensitivity analysis, indicating that our findings are trustworthy.To the best of our knowledge, several meta-analyses have reported an association between the MCP-1 rs1024611 polymorphism and DN or T2DM risk –32. For Our meta-analysis was based on thirteen studies with 2,363 patients with T2DM and 4,650 healthy controls evaluating the association between the MCP-1 rs1024611 polymorphism and T2DM risk. Ethnicity-stratified analysis indicated that the GG + GA genotype might be a protective factor against T2DM susceptibility in Caucasians but not in Asians, which agrees with a previous meta-analysis . The resOverall, our results indicate that the MCP-1 rs1024611 polymorphism has varying effects on T2DM and DN susceptibility in different ethnic groups. T2DM and DN are multietiology diseases that are genetically heterogeneous among different populations , 40. ResThis meta-analysis included eight studies with 1,074 patients with DN and 1,361 patients with T2DM and no significant risks were found in the Asian or Caucasian populations for any genetic models. This suggests that the MCP-1 rs1024611 polymorphism does not affect DN progression in T2DM. Because of the small sample size, this correlation needs to be verified by further multiethnic, large-sample-size studies. Meta-regression revealed that genotyping method was a major driver of heterogeneity in five genetic models when comparing the DN and T2DM groups. Additionally, there was a significant reduction in heterogeneity in subgroup analysis for genotyping method. However, study population, genotyping method, comorbid chronic disease, and age- and sex-adjustments were not causes of heterogeneity when comparing the DN or T2DM groups to the healthy group. Several factors significantly impacted heterogeneity (such as sex and disease stage), although the original articles did not provide more information on this. To the best of our knowledge, the present study is the first to identify a genotyping method as a source of heterogeneity in the relationship between the rs1024611 polymorphism and T2DM/DN susceptibility. To carry out a high-sensitivity rs1024611 SNP analysis, using ARMS-PCR is necessary.Meta-analysis has greater statistical power than individual studies and can produce more reliable results , 43. We In summary, the results of our meta-analysis demonstrate that the MCP-1 rs1024611 polymorphism is associated with T2DM susceptibility in Caucasians and with DN in Asians. Larger, well-designed cohort studies should be carried out in the future to verify this association.Additional file 1: Figure S1. Forest plot of T2DM risk with the dominant model (GG+GA vs. AA)  of the MCP-1 rs1024611 polymorphism. (A) for the overall populations and (B) genotyping method; (C) age- and sex- adjusted; (D) and comorbid chronic disease subgroups. Figure S2. Forest plot of DN risk with the dominant model (GG+GA vs. AA)  of the MCP-1 rs1024611 polymorphism. (A) for the overall populations and (B) genotyping method; (C) age- and sex- adjusted; (D) and comorbid chronic disease subgroups. Figure S3. Forest plot of DN risk with the dominant model (GG+GA vs. AA) (DN vs. T2DM) of the MCP-1 rs1024611 polymorphism. (A) for the overall populations and (B) genotyping method; (C) age- and sex- adjusted; (D) and comorbid chronic disease subgroups. Figure S4. Sensitivity analysis via deletion of each individual study . (A) for GG+GA vs. AA and (B) GG vs.GA + AA; (C) GG vs. AA; (D) GG vs. GA; (E) and G vs. A models. Figure S5. Sensitivity analysis via deletion of each individual study . (A) for GG+GA vs. AA and (B) GG vs.GA + AA; (C) GG vs. AA; (D) GG vs. GA; (E) and G vs. A models. Figure S6. Sensitivity analysis via deletion of each individual study (DN vs. T2DM). (A) for GG+GA vs. AA and (B) GG vs.GA + AA; (C) GG vs. AA; (D) GG vs. GA; (E) and G vs. A models. Table S1. The comprehensive search strategies for different databases. Table S2. Main characteristic of included observational studies evaluating the relationship between the MCP-1 rs1024611 polymorphism and DN/T2DM risk. Table S3. Meta-analysis of the association between the MCP-1 rs1024611polymorphism and T2DM risk . Table S4. Meta-analysis of the association between the MCP-1 rs1024611polymorphism and DN risk . Table S5. Meta-analysis of the association between the MCP-1 rs1024611polymorphism and DN risk (DN vs. T2DM).


Also, the results of this study showed that each unit increase in DII as a continuous variable generally elicited a 10% increase in the odds of E‐CRP . Subgroup meta‐analyses showed that there is a higher E‐CRP odds ratio for the articles that reported energy‐adjusted DII (E‐DII) instead of DII, the studies that measured CRP instead of hs‐CRP, and the studies that used 24‐h recall instead of FFQ as the instrument of dietary intake data collection. Individuals with a higher DII were estimated to have higher chances of developing elevated serum CRP. This value was influenced by factors such as the participants' nationality, instruments of data collection, methods used to measure inflammatory biomarkers, study design, and data adjustments. However, future well‐designed studies can help provide a more comprehensive understanding of the inflammatory properties of diet and inflammatory serum biomarkers.Diet can affect the inflammatory state of the body. Accordingly, the dietary inflammatory index (DII) has been developed to quantify the inflammatory properties of food items. This study sought to investigate the association between dietary inflammation index (DII) and the odds ratio of elevated CRP (E‐CRP) through a systematic review and meta‐analysis study. The International electronic databases of PubMed, Web of Science (ISI), and Scopus were searched until May 2023 to find related articles. From 719 studies found in the initial search, 14 studies, with a total sample size of 59,941 individuals, were included in the meta‐analysis. The calculated pooled odds ratio (OR) of E‐CRP in the highest DII category was 1.39  and the odds of elevated CRP (E‐CRP). The findings suggest that higher DII scores are associated with increased odds of E‐CRP, indicating a potential link between dietary inflammation and elevated serum CRP levels. Further well‐designed studies are needed to explore the intricate relationship between diet, inflammatory biomarkers, and their impact on health outcomes. Moreover, the continuous release of inflammatory mediators in the serum can lead to tissue damage  with six inflammatory biomarkers including IL‐1β, IL‐4, IL‐6, IL‐10, TNF‐α, and CRP, through a comprehensive literature review of the studies published from 1950 to 2010. In fact, this index is the sum of the positive and negative scores attributed to each of the aforementioned dietary components indicating their pro‐inflammatory or anti‐inflammatory potentials, respectively. A higher DII score represents a more inflammatory diet  was developed based on the association between 45 dietary components and food items . Therefore, in the present study, we aimed to assess the odds ratio of E‐CRP in relation to the dietary inflammation index in different population subgroups created based on gender, nationality, study design and methods of measuring laboratory biomarkers, data collection, and variable adjustments.22.1The current study was performed based on MOOSE Guidelines for Meta‐Analyses and Systematic Reviews of Observational Studies  observational studies with cohort, case–control, or cross‐sectional designs; (2) Investigated the relationship between DII and CRP as the outcome variable in a wide range of clinical settings; (3) performed on a population of adults (≥18 years); (4) reported odds ratios [ORs], risk ratios [RRs], or hazard ratios [HRs] along with their 95% confidence intervals [CIs] for the association between DII and CRP. If findings from one study were published in more than one article and/or database, we selected the most credible version. We excluded letters, comments, short communications, review articles, and ecological and animal studies.2.3Required data from each eligible study were extracted by two independent investigators (MH and SS), and any disagreements were reconciled by discussion or referral to a third reviewer (EY). Any reported ORs or HRs or RRs and corresponding 95% CIs for the association between DII and CRP were extracted from each study. In addition to effect sizes (ESs), the following information was extracted: first author's name, year of publication, country of origin, demographic characteristics of participants (age range and gender), number of participants and cases, duration of follow‐up for prospective studies, methods used for exposure and outcome assessment, and confounding variables adjusted in the statistical analysis. All extracted data were included in a standardized Microsoft Excel.2.4Q‐test and I2 test. I2 values of >50% were considered significant between‐study heterogeneity. Publication bias was examined using funnel plot visual inspection and Egger's regression asymmetry tests. A trim‐and‐fill method was used to detect the effect of probable missing studies on the overall effect. All statistical analyses were performed using Stata version 14.0 . All p‐values were two‐sided, and p < .05 was considered statistically significant.For comparison of the highest versus lowest categories of DII, ORs and RRs  were used to calculate weighted ORs and RRs. To estimate the pooled effect size, random effects model was used for analyses. Existence of heterogeneity was checked by Cochran's 2.5As illustrated in Table 3During the initial step of our meta‐analysis search, a total of 719 publications were identified. Among these, 395 articles were removed due to duplication, and an additional 123 studies were excluded based on their titles and abstracts as they were deemed irrelevant. Following careful screening, 201 publications were selected for full‐text eligibility assessment. After a thorough evaluation, we included 14 clinical trials in this systematic review and meta‐analysis, meeting our predefined inclusion criteria  in eight papers, providing a comprehensive overview of participants' long‐term dietary habits. Additionally, 24‐h dietary recall (24HR) was employed in five articles, offering a snapshot of participants' dietary intake within a specific 24‐h period. Further details regarding the main characteristics of the included studies can be found in Table 3.1p = .63, and I2 = .0%) in comparison with the lowest DII category    of E‐CRP in the highest DII category was 1.39 . However, we observed a significant publication bias in studies that examined DII as a categorical variable (p = .005) but the application of the trim‐and‐fill method did not alter the pooled effect size, indicating that the results were not affected by the publication bias. Funnel plots for visual test of publication bias are also provided in Figure Egger's regression test showed that publication bias was not significant for studies that examined the DII variable as a continuous independent variable p = .. Howevere p = .00 but the 4This study showed that higher DII scores were associated with higher odds of having elevated plasma CRP levels in a population of men and women of different nationalities. Thus, our results suggest that adherence to pro‐inflammatory diet that covers low fruits and vegetables, high SFA, and high refined carbohydrates is associated with higher levels of the blood inflammatory biomarker CRP. The findings of this study show that dietary inflammation index plays an important role in changing the inflammatory status of the body, which is congruent with some previous studies, where it has been shown that this index is associated with several health consequences, such as mortality, cardiovascular disease, and kidney disease  has been shown to be lower than in the western male population, which is reported to range from 2 to 3 mg/L , it could be a source of heterogeneity in our findings and should be verified in future studies. These methods are used to determine suspended solids in liquids by measuring the light intensity before and after exposure to the sample. In Nephelometry, the intensity of scattered light is measured, while in Turbidometry, the intensity of light passing through the sample is measured  vs. FFQ to collect dietary data. This may be due to the different properties of these two data collection instruments. There is evidence that information obtained from 24‐h recall may be more accurate (less biased) estimates than that from FFQ in epidemiological studies  . There were variations in the number of food parameters used to calculate the DII across the included studies, which may have introduced heterogeneity in the results. Moreover, the studies included in the present meta‐analysis had been conducted in a limited number of countries in North America, Europe, and East Asia. Therefore, generalizing the results of this study to the global community is unfeasible, because the eating habits of people from different societies are highly variable under the influence of their cultural and religious beliefs. Additionally, it is important to note that the study protocol was not registered in a trial registry such as PROSPERO, which could have enhanced transparency and reduced potential bias. However, we took measures to ensure rigorous methodology and adherence to established guidelines, such as the MOOSE Guidelines for Meta‐Analyses and Systematic Reviews of Observational Studies.Despite the aforementioned limitations, there are numerous strengths worth noting. We performed subgroup meta‐analyses based on possible sources of heterogeneity, including the methods of dietary and laboratory data assessment, studies' country of origin, and adjustment of important confounding variables. Also, we performed meta‐regression to examine the possible association between the findings with the sample size and the year of publication of the included studies. However, we had to exclude some studies from the subgroup meta‐analyses due to insufficient information.5We found in the present study that the odds ratio of E‐CRP was higher in the population with higher dietary inflammation index. Men in the highest category of DII were more likely to experience E‐CRP, whilst this value was insignificant in the female population. Also, the odds ratio of E‐CRP, associated with DII, was under the influence of factors such as the nationality of the study population, methods of dietary and laboratory data assessment, study design, and adjustment of BMI and energy. Given that the studies included in the present meta‐analysis were originally from specific regions of the world, it is suggested that well‐designed studies be conducted in the future to provide a more comprehensive insight into the issue of the correlation between diet and inflammatory serum biomarkers.Somaye Saboori: Conceptualization (lead); project administration (lead); writing – review and editing . Salman Mohammadi: Formal analysis (lead); writing – original draft . Mahboobe Hosseinikia: Data curation ; investigation ; writing – review and editing . Ali Ghaffarian‐Bahraman: Investigation ; writing – original draft . Cain C. T. Clark: Validation ; writing – review and editing . Ian Davies: Validation ; writing – review and editing . Esmaeil Yousefi Rad: Data curation ; investigation ; writing – review and editing .The current study was funded by Lorestan University of Medical Sciences, Khorramabad, Iran. The funder has played no role in the design of the study or in the collection, analysis, or interpretation of the data.The authors declare that they have no conflict of interest.Table S1Click here for additional data file.Table S2Click here for additional data file.


Different studies provide conflicting evidence regarding the potential for glucocorticoids (GCs) to increase the risk of cardiovascular diseases. This study performed a systematic review and meta-analysis to determine the correlation between GCs and cardiovascular risk, including major adverse cardiovascular events (MACE), death from any cause, coronary heart disease (CHD), heart failure (HF), and stroke.We performed a comprehensive search in PubMed and Embase . Studies that reported relative risk (RR) estimates with 95% confidence intervals (CIs) for the associations of interest were included.A total of 43 studies with 15,572,512 subjects were included. Patients taking GCs had a higher risk of MACE , CHD , and HF . The MACE risk increased by 10% (95% CI: 6%–15%) for each additional gram of GCs cumulative dose or by 63% (95% CI: 46%–83%) for an additional 10 μg daily dose. The subgroup analysis suggested that not inhaled GCs and current GCs use were associated with increasing MACE risk. Similarly, GCs were linked to an increase in absolute MACE risk of 13.94 (95% CI: 10.29–17.58) cases per 1,000 person-years.Administration of GCs is possibly related with increased risk for MACE, CHD, and HF but not increased all-cause death or stroke. Furthermore, it seems that the risk of MACE increased with increasing cumulative or daily dose of GCs. Since their discovery in the 1930s, glucocorticoids (GCs) are an essential class of anti-inflammatory and immunosuppressive drug widely used for treating medical conditions such as asthma and chronic obstructive pulmonary disease, rheumatoid arthritis, inflammatory bowel disease, and lymphoid malignancies . It has With these inconsistent findings in mind, the cardiac safety of individual GCs required further investigation to guide clinical decisions better. Therefore, we performed a systematic review and meta-analysis of published studies to comprehensively determine the correlation between GCs and CVDs, including major adverse cardiovascular events (MACE), death from any cause, coronary heart disease (CHD), heart failure (HF), and stroke. We tested the hypothesis that an increased dose of GCs use is related to a greater risk of MACE.For the meta-analysis, the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) checklist was used . The traTo minimize differences between studies, the inclusion criteria were as follows: (1) studies presented the use of GCs (users and non-users), including cohort, case-control studies, and randomized controlled trials; (2) studies reported outcomes of CVDs with the minimum information necessary to estimate the relative risk (RR). Moreover, we excluded review articles, letters to editors, and case reports. The decision for inclusion of each study was made independently by two authors. Conflicts between reviewers were resolved by consensus.Two authors independently extracted relevant data such as the year of publication, geographical location, source of study, study design, baseline patient characteristics, the definition of exposure, the outcome of CVDs, and the cumulative and daily dose of GCs use. We assessed the quality of observational studies using the Newcastlee-Ottawa scale . We asseThe primary outcome of this study was MACE, including CHD, stroke, HF, and cardiovascular death. Additionally, we examined CHD, HF, and stroke all-cause death as secondary outcomes.The primary measure of the association between GCs and CVDs in each study was relative risks (RR). Because the incidence of CVDs is rare, hazard ratio (HR) and odds ratio (OR) were treated as equivalent estimates of RR –18. We eRandom effects were used to calculate pooled RRs (95% CI) to account for heterogeneity among studies, and the results are treated as conservative estimates because CIs are wider. Cochran’s Q and I2 statistics were used to quantify heterogeneity across estimates, with an I2 greater than 50% indicating significant heterogeneity . Meta-reIn the dose-response analysis, the generalized least squares (GLST) regression model and restricted cubic splines were used to assess pooled dose-response relation between GCs and MACE. Both linear models and nonlinear models were fitted, and the results presented 95% CIs , 23. Whi0]. RR indicates pooled RRs and I0 was the cumulative incidence of events among patients without GCs * Ihout GCs .p-value <0.05 was considered statistically significant.All statistical analyses were performed using STATA 15.0 . A two-sided As displayed in The studies included 33 population-based cohort studies, nine case-control studies, and one randomized controlled clinical trial. Among the included studies, five were from Asia, 16 from Europe, and 22 from North America. Data for cumulative and daily GCs doses were available in 7 and 11 studies, respectively. The main characteristics, including references to the studies, were presented in As shown in Q = 190.64, I2 = 88.5%, P < 0.001). Therefore, subgroup analysis and meta-regression were performed in different subgroups across various main study characteristics and clinical factors to identify potential sources of heterogeneity (P = 0.686), and Begg test (P = 0.751) revealed no evidence of publication bias.Twenty-three studies  with 883,877 cases reported the outcome of MACE. Overall, GCs use was significantly associated with increased MACE risk compared with no use , and the dosage was converted into a prednisolone-equivalent dose . As depiQ = 111.56, I2 = 83.0%, P < 0.001), and the type of GCs, male proportion, and publication year showed potential sources of between-study heterogeneity (P = 0.586) and Begg (P = 0.922) tests showed publication bias.For the risk of CHD, 20 studies  contained information on GCs use, and GCs were associated with an increased risk of CHD . In studies included, the Yao’s study contains over 14 million patients, which represents over 90% of the subjects studied. Thus a separate analysis excluding this study was performed, and we also found that GCs were associated with an increased risk of HF . In the subgroup analysis, we found the differences in the type of study, age, and time periods of GCs use would contribute to heterogeneity .Seven studies about the relationship between GCs and HF contain data from 14,811,957 patients. A significantly increased risk of HF was observed, for GCs use . Differences in study type, events, GCs type, age, male proportion, time periods of GCs use, whether adjusted for risk profiles, and publication year were not potential sources of heterogeneity (P = 0.671), and Begg (P = 0.827) suggested an absence of publication bias.ogeneity . Funnel ogeneity , Egger . We also conducted a subgroup analysis and meta-regression in different subgroups  . AdditioContrary to expectations, this study did not find a significant difference between GCs and stroke, although there were shared many risk factors contributing to the occurrence of both stroke and CHD. It is well established that GCs were available to act on the vascular wall, endothelial, and vascular smooth muscle . Also, bIn addition, this study’s observed difference between GCs and all-cause death was insignificant. It could be due to the GCs’ anti-inflammatory effect being favorable for some participants’ survival though GCs increased MACE risk. In a previous study, The Lung Health Study Research Group demonstrated that corticosteroids might attenuate airway hyperresponsiveness , which wThe results suggested that the association of GCs with MACE risk may be mediated by an acute mechanism for the higher risk in current compared to ever GCs users. Some plausible explanations have been proposed. First, Börcsök et al. reported that plasma endothelin-1 (a vasoconstrictor associated with cardiovascular disease) level increased 50% early in human subjects after prednisolone use . Second,Notably, based on the subgroup analysis, inhaled GCs were associated with a reduction in MACE and all-cause death without increasing the risk of CHD or stroke. The mechanisms that underlie these responses, however, are poorly defined. As mentioned above, the risk of MACE increased with increasing GCs cumulative dose. It is possible that harmful systemic reactions were avoided because very little of GCs could become systemic when GCs were administered topically, bringing substantial benefit. Sin et al. reported that ICS therapy reduced C-reactive protein (CRP) levels by 50% after two weeks . Thus, IOur findings have important medical implications. It remains unclear whether GCs cause adverse cardiac events, and the risks of GC use have attracted much attention among clinicians and patients. Our research timely evaluates the evidence of suspected risks. The increased risk of CVDs, including MACE, CHD, and HF, was observed in our study, which supported close monitoring of CVDs is warranted in patients during GCs treatment. Also, clinicians should be mindful that regular use of high-dose GCs should be avoided where possible, and decrement and discontinuance in GCs should be early. In addition, inhaled GCs seem to be safe in patients.Our study has several strengths, including the strict criteria of inclusion, the large sample size, the diversity of the study population, the dose-response relationship between GCs and MACE, and pre-specified subgroup analyses. Besides, the robustness of the study findings is supported by the absence of important publication bias.Nevertheless, several limitations of our study were also discussed. First, the study has its inherent limitations of being a retrospective analysis. Second, there is substantial heterogeneity between the different studies. The therapeutic protocols of GCs were not equivalent in different studies. Population characteristics, including age, sex, presence or absence of hypertension, diabetes, and dyslipidaemia might also introduce heterogeneity into studies. And as only one study in this literature search is a randomized-controlled trial, the data are susceptible to indication bias. Third, adjusted hazard ratios were not used in some included studies, which may also influence outcomes. Finally, comprehensive individual patient data were unavailable from the studies, we were thus unable to identify the independent associations of individual variables with study outcomes. Such as, the conclusions regarding GCs dosing would be influenced as it is not clear if the GCs were dosed based on weight, severity of illness , clinical guidelines/local protocols and this will likely be impossible to fully investigate with this meta-analysis approach. Furthermore, the lack of individual patient data limited further analysis such as meta-regression.In conclusion, the results of this meta-analysis reveal that GCs is possibly related with increased risk for MACE, CHD, and HF but not all-cause death or stroke. The dose-response analysis highlighted the risk of MACE may increased with increasing GCs cumulative or daily dose was observed. In addition, it seems that not inhaled GCs, and current GCs use were associated with higher risk. The association between GCs and CVDs has clinical relevance with respect to individual screening and the prevention of CVDs. This calls for further large RCTs warranted to confirm these findings and, additional studies will be needed to clarify the underlying mechanism.


Despite growing evidence for the association of adherence to the Mediterranean diet with gastric cancer risk, the results remain inconclusive. The purpose of this systematic review and meta-analysis was to summarize the evidence from previous observational studies and assess the potential association between adherence to the Mediterranean diet and risk of gastric cancer using a dose–response meta-analysis.Q test and I-squared (2I) statistic were used to detect the sources of heterogeneity among the included studies.A comprehensive literature search for all observational studies published up to June 30, 2023 was conducted using the databases of PubMed, ISI Web of Science, EBSCO, China National Knowledge Infrastructure (CNKI) and Wanfang Data. The pooled relative risks (RRs) and 95% confidence intervals (CIs) were calculated for the highest versus the lowest categories of Mediterranean diet score in relation to gastric cancer risk, using random-effects models. The Cochran’s p < 0.001). In addition, linear dose–response analysis showed that each 1-score increment in Mediterranean diet score was associated with a 5% lower risk of gastric cancer . Stratified analysis showed a significant association between adherence to the Mediterranean diet and risk of gastric cancer in case–control studies , and a marginally significant association in prospective cohort studies , respectively. At the same time, a more significant association between Mediterranean diet and reduced risk of gastric cancer was observed in other countries  than in Western countries .Overall, 11 studies (five cohort and six case–control studies) with a total number of 1,366,318 participants were included in the final analysis. Combining 14 effect sizes from 11 studies revealed that compared with the lowest category, the highest adherence to the Mediterranean diet was associated with a 29% reduction in the risk of gastric cancer (RR:0.71; 95%CI:0.59–0.84, Our results demonstrate that high adherence to the Mediterranean diet is associated with 29% reduced risk of gastric cancer. Further large prospective studies and randomized controlled trials are warranted to confirm our findings. Helicobacter pylori (H. pylori) infection, cigarette smoking, alcohol intake, and dietary factors  statement . MoreoveWe carried out a comprehensive search of articles published up to June 30, 2023 using PubMed, ISI Web of Science, EBSCO, CNKI and Wanfang Data with no restrictions in language or publication date. The following terms were used in our search: { AND }. In addition, a manual search in the reference lists from the retrieved articles or reviews or meta-analyses was performed to find the potentially eligible studies. All of these steps were accomplished by two independent reviewers (ZQ and SL), and any disagreements with article selection were resolved through discussion with Y-LF. Our selection criteria was based on the PECOS  framework, which is shown in in vitro studies; (4) studies not reported as HRs, RRs or ORs with 95%CIs.Two independent reviewers (QZ and LS) carried out an initial screening of all titles and abstracts from retrieved articles to identify eligible studies that should be included in the analysis. Any disagreements were settled by discussion or in consultation with the third reviewer (Y-LF). When all reviewers agreed, the full-text articles were reviewed against inclusion and exclusion criteria for the present systematic review and updated meta-analysis. To be included in the present meta- analysis, articles had to meet the following criteria: (1) observational studies  conducted in adult population (aged ≥18 years); (2)considered adherence to the Mediterranean diet as the exposure; (3) assessed the association between adherence to the Mediterranean diet and risk of gastric cancer; (4) provided estimates of RRs, HRs, ORs with their corresponding 95%CIs; (5) If the data in retrieved article lacked sufficient detail, the corresponding author of the original study was contacted by email; (6) gastric cancer diagnoses were confirmed by clinical interviews, or self-report on a previous physician-made diagnosis of gastric cancer. Moreover, studies were excluded if they fulfilled one of the following criteria: (1) unrelated articles; (2) non-observational studies, e.g., reviews, editorials, case reports and conference letters; (3) animal, cell culture, and After completing selection of all eligible studies, two independent authors (LS and FZ) extracted the following information: first author’s last name, publication year, study design, country, sample size, number of gastric cancer cases, mean age/age range, follow-up time (cohort studies), components of the Mediterranean diet score, methods of dietary assessment, reported risk estimates (HR/OR/RR) and the corresponding 95%CIs and confounding factors that were adjusted for in the multivariate analyses. Any discrepancies and disagreements about data extraction were resolved by consensus or discussion with the third author (Y-LF).The Newcastle-Ottawa Scale (NOS) was used to evaluate the overall quality of the included studies in the present study . We assi0) + (P0*OR)], in which P0 indicates the incidence of the outcome of interest in the non-exposed group  was used to pool the RRs and 95%CIs of the highest versus the lowest category of Mediterranean diet in relation to gastric cancer. Otherwise, the fixed effect model is used . A p-value less than 0.05 (two-tailed) was considered statistically significant.The reported HRs in the primary studies were considered as equal as RRs . ORs wered group . Log-tra is used . If sign is used . If ther is used . Finallyp < 0.001). There was evidence of substantial heterogeneity between studies , and therefore the effect was assessed using the random-effects model.Eleven articles comprising 5,708 gastric cancer cases and 1,366,318 participants, were included to assess the link between adherence to the Mediterranean diet and the risk of gastric cancer. Combining 14 effect sizes from 11 studies, p < 0.001; Pnon-linearity = 0.330)    . The ana= 0.382) . In addi= 0.538) .I2 = 79.7%; p < 0.001), subgroup analyses were performed to explore the potential sources of heterogeneity (p = 0.024), with less evidence of heterogeneity between studies . In contrast, there was also significant association between Mediterranean diet and decreased risk of gastric cancer in case–control studies , with evidence of heterogeneity . For study area, we found a significant inverse association between adherence to Mediterranean diet and gastric cancer risk in other countries . There was no evidence of heterogeneity between studies . When the results were stratified by mean age, we found a significant inverse association between Mediterranean diet and gastric cancer risk in age ≥ 50 and age < 50 . The heterogeneity was most apparent in age ≥ 50 . For sample size, we found a significant inverse relationship between Mediterranean diet and gastric cancer risk in the subgroups of sample size > 5,000 . However, the heterogeneity was less . In addition, a protective association was found between Mediterranean diet and gastric cancer risk in the studies with sample size<5,000 , and there was significant heterogeneity . We also performed stratified analysis based on sex, and results showed the inverse association between Mediterranean diet and gastric cancer risk in male and female (0.77 vs. 0.91). however, there was no evidence of heterogeneity in female .Given the high heterogeneity of this study . Due to no evidence of publication bias, we could not perform the trim and fill analysis to adjust the pooled effect estimates. All included studies in our analyses have received a NOS score ≥ 7, and these studies were regarded to be of high quality  .To our knowledge, this is the latest systematic review and meta-analysis to evaluate the association between adherence to the Mediterranean diet and gastric cancer risk with 1,366,318 participants and 5,708 cases of gastric cancer. The current systematic review and meta-analysis of 11 epidemiological studies shows that greater adherence to the Mediterranean diet was significantly associated with a 29% reduction in the risk of gastric cancer. In addition, the dose–response analysis also shows that each 2-score increment in Mediterranean diet was associated with a 6% reduction in the risk of gastric cancer. In a meantime, sensitivity analysis revealed that the summary effect of Mediterranean diet on gastric cancer was not substantially modified by excluding a certain study. Collectively, the results of this systematic review and meta-analysis provide further scientific evidence supporting the adoption of adherence to the Mediterranean diet for the primary prevention of gastric cancer.a posteriori methods to evaluate the association between certain dietary patterns  and gastric cancer . As far as we know, inter-study heterogeneity is common in previous meta-analyses  declare financial support was received for the research, authorship, and/or publication of this article. This study was supported by the National Natural Science Foundation of China (grant number: 82004040), Medical and Health research fund project of Zhejiang Province (no. 2022KY006) and Traditional Chinese Medicine Research Project of Zhejiang . The sponsors played no role in the study design, data collection, or analysis, or decision to submit the article for publication.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.


Periodontal disease is associated with many systemic diseases, such as cardiovascular diseases, atherosclerosis, diabetes mellitus, stroke, and pulmonary diseases. Interestingly, recent literature suggests that periodontal disease might be a risk factor for various cancers such as lung, colon, oesophageal, head, and neck cancers. However, the precise mechanistic link is lacking. Hence, in this meta-analysis, we aimed to investigate the correlation between periodontal disease and lung cancer in periodontally diseased patients. Data were searched for relevant studies from 2010 to 2022. We correlated periodontal disease and lung cancer based on adjusted ORs/HRs and associated CIs. I2 statistic was used to assess statistical heterogeneity. Publication bias was analyzed by visually inspecting the symmetry of the funnel plot and Egger's test. The study is registered in the International Prospective Register of Systematic Reviews . A total of 194,850 participants from observational studies (two case-control studies and five cohort studies) were incorporated for the current analysis. The meta-analysis of included studies showed an overall effect size (risk ratio) of the periodontal disease group with respect to the non-periodontal disease group for lung neoplasm to be 1.41 (95% CI: 1.32-1.52). The value was more than 1, indicating that the periodontal disease group had a relatively higher lung cancer prevalence than the non-periodontal disease group. Further, the overall risk ratio was found to be statistically significant (p<0.00001). Moreover, the funnel plot suggested some degree of publication bias. Evidence in our study implicated that there is an increased risk of occurrence of lung cancer in chronic periodontitis patients. Cancer is a significant etiologic factor in lowering life expectancy and one of the main reasons for death worldwide . In 2020MethodologyStudy SelectionThis systematic review and meta-analysis were done to determine the risk of lung cancer in chronic periodontitis patients. Three hundred twenty-five studies were initially obtained from the literature search . Initially, 15 studies were included based on the titles and abstracts of the articles. Out of these, seven studies were assessed for our meta-analysis, comprising five cohort studies and two case-control studies. The criteria for study selection were based on Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA)  was used to assess the method's quality, including participant selection, comparability, result assessment, and exposure Table  8]..8].Search StrategyWe explored the observational studies demonstrating a correlation between lung cancer and periodontal disease through a search engine from January 2010 to December 2022. All related studies in English were identified with the following search terms "periodontal disease" OR "periodontitis" OR "probing pocket depth" OR "pyorrhoea" AND "lung cancer" OR "neoplasm of lung."Eligibility CriteriaScreening of the title and abstract of all studies was done. The complete text of seven likely pertinent articles -15 was eData Collection ProcessThe risk of bias/quality assessment of all the articles included was done by two reviewers (XYZ and MYX) who assessed the studies based on inclusion and exclusion criteria. The author's surname, study duration, year of publication, sample size, study design, population characteristics, exposure, outcome, and adjusted data of HRs/ORs/RRs and 95% CI were assessed. Information based on the above criteria was discussed between the authors for any discrepancies.Data AnalysisRevMan (version 5), a Cochrane software, was used for preparing the review and performing a meta-analysis of our data. The adjusted ORs/HRs and their associated CIs were used as a measure to correlate between lung cancer and periodontal disease, and data was extracted from included studies. The risk ratio of the periodontal disease group with respect to the non-disease group was used to measure the dichotomous outcomes and effect size, both with 95% CI. I2 statistics were used to determine the heterogeneity of the data across studies  of the periodontal disease group with respect to the non-periodontal disease group for lung neoplasm was 1.41 (95% CI: 1.32-1.52)  in older women, irrespective of smoking, during 8.32 years of mean follow-up . In anotThe study found conflicting evidence for a link between periodontal disease and mortality from all causes, all cancers, and specific cancers like lung cancer. Risk factors for mortality were being a man, having less education, and smoking . Apart fPeriodontal disease is an inflammatory condition which is caused by a periodontal pathogen and often accompanied by the elevated level of biomarkers like interleukins , interferons (IFN-ɣ) and C-reactive protein (CRP) -22. It iLimitations of the studyThere are certain limitations in our meta-analysis. Firstly, studies that are included are few and the study was conducted in different regions of the world so there can be a variation in the incidence of lung cancer in periodontitis patients. Secondly, there are some confounder factors such as smoking, diabetes, and alcohol which should always be kept in mind for proper assessment of the relationship between periodontitis and lung cancer. Further treatment modalities should be incorporated to predict the reduced risk of cancer in a chronic periodontitis-treated patient. Additionally, there was a lack of data regarding the specific association between clinical parameters of periodontitis and specific histological types of lung cancer. Furthermore, there was a major difference in reporting clinical parameters methods for chronic periodontitis assessment and lung cancer association between included studies which might create a source of heterogeneity.The results of our meta-analysis summarized and suggested a possible correlation between periodontitis and lung cancer risk. Chronic periodontitis is an inflammatory condition of the periodontium. It can elevate the risk of lung cancer in susceptible individuals by creating an inflammatory milieu. Preventive and therapeutic measures are well known for periodontal diseases and should be implemented, keeping in mind improving the local as well as overall systemic health of the patient. However, to further support this association between the two conditions, large-scale studies are required with adequate control/adjustment of confounding factors for a more conclusive result.


Although trimethylamine-N-oxide (TMAO) shows a notable correlation with cardiovascular disease, its association with acute ischemic stroke (AIS) remains uncertain and necessitates further investigation.A meta-analysis was conducted to assess the relationship between trimethylamine-N-oxide and acute ischemic stroke.We conducted a comprehensive search in PubMed, Embase, Cochrane, CNKI, VIP, Wanfang, and CBM, spanning from their inception to 23 September 2023. The search was consistently updated and supplemented by bibliographies of retrieved articles and previous reviews. A total of 20 eligible studies, including 17 case‒controls and 3 cohort studies, were selected, involving 9141 participants . For the dose‒response analysis, three case–control studies were eligible. We extracted and pooled TMAO mean and standard deviation from observational studies for control and ischemic stroke groups. The effect sizes were combined using the random-effects model. Where possible, dose‒response analysis was performed.2 = 1.6%, P = 0.4484).Overall, the pooled standardized mean difference (SMD) demonstrated significantly higher concentrations of serum/plasma TMAO in AIS compared to the control group . Additionally, the dose‒response meta-analysis revealed a 12.1% relative increase in the risk of acute ischemic stroke per 1 μmol/L rise in TMAO concentration (RR = 1.12; 95% CI 1.07–1.17; P<0.05; IThese findings indicate a potential increased risk of AIS associated with elevated TMAO levels. Stroke is a significant global public health issue, being a major contributor to both mortality and disability , 2. ClinGut microbes metabolize choline-rich foods to generate trimethylamine (TMA), subsequently converted to trimethylamine-N-oxide (TMAO) in the liver by flavin monooxygenase family members, FMO1 and FMO3 , 6. IndeWe conducted a comprehensive literature search in PubMed, Embase, VIP, CNKI, CBM, Wanfang Database, and Cochrane Library, covering the period from inception to 23 September 2023. The search strategy employed the following terms: (1) Trimethyloxamine, including Medical Subject Headings (MeSH) terms trimethylammonium oxide, trimethylamine N-oxide, TMAO, and trimethylamine oxide; (2) Ischemic Stroke, encompassing MeSH terms ischemic strokes, Stroke, Ischemic, Ischaemic Stroke, Ischaemic Strokes, Stroke, Ischaemic, Cryptogenic Ischemic Stroke, Cryptogenic Ischemic Strokes, Ischemic Stroke, Cryptogenic, Stroke, Cryptogenic Ischemic, Cryptogenic Stroke, Cryptogenic Strokes, Stroke, Cryptogenic, Cryptogenic Embolism Stroke, Cryptogenic Embolism Strokes, Embolism Stroke, Cryptogenic, Stroke, Cryptogenic, Embolism, Wake-up Stroke, Stroke, Wake-up, Wake up Stroke, Wake-up Strokes, Acute Ischemic Stroke, Acute Ischemic Strokes, Ischemic Stroke, Acute, Stroke, and Acute Ischemic; and (3) Stroke, including MeSH terms Strokes, Cerebrovascular Accident, Cerebrovascular Accidents, CVA (Cerebrovascular Accident), CVAs, Cerebrovascular Accident, Cerebrovascular Apoplexy, Apoplexy, Cerebrovascular, Vascular Accident, Brain, Brain Vascular Accident, Brain Vascular Accidents, Vascular Accidents, Brain, Cerebrovascular Stroke, Cerebrovascular Strokes, Stroke, Cerebrovascular, Strokes, Cerebrovascular, Apoplexy, Cerebral Stroke, Cerebral Strokes, Stroke, Cerebral, Strokes, Cerebral, Stroke, Acute, Acute Stroke, Acute Strokes, Strokes, Acute, Cerebrovascular Accident, Acute, Acute Cerebrovascular Accident, Acute Cerebrovascular Accidents, Cerebrovascular Accidents, and Acute. These terms were combined as follows: 1 and (2 or 3). The search history is available in the The inclusion criteria for the meta-analysis were as follows: (1) Study participants were adults aged ≥ 18. (2) The exposure factor was TMAO, and studies reported either the mean and standard deviation of TMAO for the two groups or the odds ratio with corresponding 95% confidence interval for different increments of TMAO. (3) The included studies focused on ischemic stroke patients, adhering to internationally accepted guidelines for defining and describing acute ischemic stroke. (4) The selected articles were observational studies, including cross-sectional, case‒control, and cohort studies. Exclusion criteria were as follows: (1) Retrieved articles included reviews, review articles, conference papers, letters to editors, case reports, meta-analyses, animal experiments, or unrelated studies. (2) Articles lacking mean and standard deviation data for TMAO levels in the case and control groups or relevant data to calculate them. (3) Studies that aggregated outcomes into total stroke without reporting AIS separately. (4) Insufficient availability of detailed study information.Two researchers independently evaluated the eligibility of the literature. In case of any disagreements, a third researcher was consulted, or a group discussion was held to reach a consensus. The protocol for this review was registered with PROSPERO (CRD42022373779).Following the PRISMA guidelines, two authors (Z.S. and Y.H.) extracted the following characteristics from the included studies: first author’s surname, publication time, study origin, cohort name, sample size, number of cases, age at entry, sex distribution, outcome, method of measuring TMAO, and the mean and standard deviation of TMAO for both control and AIS patients separately. All specimens were collected promptly after admission. For the dose‒response meta-analysis, risk estimates (preferably adjusted measures), hazard ratios (HRs), risk ratios (RRs), or odds ratios (ORs) with their corresponding 95% confidence intervals (CIs) were utilized, along with adjustment factors. In cases where multiple risk estimates were provided, the multivariable model was chosen. Study quality assessment was conducted using the Newcastle–Ottawa Quality Assessment Scale (NOS), which evaluates three quality parameters  divided into eight specific items. Studies scoring ≤ 3 were classified as low quality, 4–6 as moderate quality, and ≥ 7 as high quality .2 value, with values below 25% indicating low heterogeneity, 25–50% indicating moderate heterogeneity, and above 50% indicating high heterogeneity [The statistical analyses were conducted using STATA V.14 . The primary studies reported TMAO concentration in various formats, either as a continuous or categorical variable. To ensure consistency across the meta-analysis, we extracted the mean and standard deviation of TMAO for the control and AIS groups from 19 selected articles. In cases where this data was not available, we applied the transformation method proposed by Dehui Luo et al. , 16. Hetogeneity . The resogeneity . This meA meta-regression analysis was employed to investigate how significant covariates might impact the heterogeneity observed between studies. We established statistical significance by considering P values equal to or less than 0.05 . SubgrouA comprehensive computerized database search initially identified a total of 504 papers, with the search and screening process illustrated in All of the included studies obtained a rating of 6 to 8 stars on the Newcastle‒Ottawa Scale. Specifically, six studies received a score of 6, eight studies received a score of 7, and five studies received a score of 8. The agreement between the two investigators for quality assessment, as measured by the Kappa statistic, was 0.922.2 = 97.6%, P<0.01), indicating a notable difference in effect size between the AIS group and the control group using a random effects model. The analysis revealed a higher effect size in the AIS group compared to the control group  for AIS with each 1 μmol/L increment of TMAO , while it was lower in the cohort study subgroup . The effect size appeared to be lower in the cohort study subgroup , consistent with the results obtained from the subgroup analysis.Sensitivity analysis was performed by comparing the combined effect size  to assess the robustness of the results. The analysis revealed that removing any single study did not overturn the findings, indicating the reliability of the combined results see .The funnel plot exhibited asymmetry, suggesting the presence of publication bias in the included studies see . To evalThis study represents the first comprehensive systematic review and dose‒response meta-analysis investigating the relationship between circulating levels of trimethylamine N-oxide (TMAO) and the risk of acute ischemic stroke (AIS), based on a thorough search of meta-analyses listed on the PROSPERO platform. A total of 19 relevant articles were meticulously selected for inclusion in this analysis. The findings indicate a significant elevation in serum/plasma TMAO levels among AIS patients compared to the control group, suggesting the potential involvement of TMAO in the development of AIS. Furthermore, the dose‒response meta-analysis, incorporating three studies, demonstrated a consistent 12.1% relative increase in AIS risk for each additional 1 mol/L of TMAO, indicating a linear correlation between TMAO levels and the risk of AIS. Importantly, this strong association between TMAO plasma levels and AIS risk remained consistent across all tested subgroups and study populations, underscoring the robustness and generalizability of the findings. These results highlight the potential of TMAO as a valuable biomarker for identifying individuals at high risk of AIS and suggest it may also serve as a promising therapeutic target for reducing the risk of AIS.Randomized controlled trials (RCTs) are widely regarded as the gold standard for establishing causal relationships between interventions and outcomes. However, due to the limitations of available evidence, our study relied on observational studies, including case‒control and cohort studies, which are susceptible to selection bias, recall bias, and reverse causality. It is worth noting that the cohort subgroup consisted of only three articles, potentially introducing bias. The study design itself can significantly influence the correlation between TMAO levels and AIS risk. Although our meta-analysis provides evidence of an association between circulating TMAO levels and AIS risk, it is important to interpret these findings cautiously given the inherent limitations of the included studies. Regarding heterogeneity, the cohort study exhibited high heterogeneity exceeding 50%. However, the heterogeneity decreased significantly, suggesting that the study design may contribute to observed heterogeneity to some extent. Nonetheless, due to the limited number of articles included, further confirmation is required. Furthermore, sensitivity analyses indicated that the results were robust to the removal of specific regions. However, it is worth noting that only one of the included articles was conducted in the United States, where dietary patterns and sources of TMAO may differ from those in China and other regions. TMAO is primarily obtained from foods containing choline or carnitine, and variations in dietary structures may result in divergent TMAO levels. However, due to the limited number of studies in our analysis, we were unable to thoroughly explore whether the correlation between TMAO and dietary patterns would vary across regions. Notably, high-quality investigations examining the association between TMAO and cardiovascular risk have primarily been conducted in the United States and Europe. Despite potential differences in dietary patterns and lifestyle factors, these studies consistently demonstrate a positive correlation between TMAO levels and the risk of cardiovascular disease. Therefore, additional investigations encompassing diverse regions and populations are crucial to verify and strengthen our current understanding of this relationship. Although subgroup analyses and one-way regression analyses were performed to explore potential influencing factors, neither method revealed any significant factors affecting the findings. Moreover, the dose‒response analysis adjusted for various AIS-related risk factors, including blood pressure, renal function, and BMI, and the results still indicated a positive association between TMAO levels and AIS prevalence. It is important to acknowledge that publication bias was detected in our analysis. However, the cut-and-patch method was employed to verify that the bias had minimal impact on the results. Additionally, our study included both published and unpublished studies to mitigate the influence of publication bias. It is possible that negative results tend to remain unpublished, which could explain the inclusion of unpublished literature in our study.Ischemic stroke can arise from various events, such as cardiac embolism, cerebral vascular occlusion, and atherosclerosis, all affecting cerebral circulation . PreviouIt is crucial to acknowledge the limitations of this study. The majority of included studies were case‒control studies, with a limited number of cohort studies and no randomized controlled trials. Further research, particularly RCTs, will be necessary to establish a causal relationship between TMAO and AIS risk. Additionally, future studies should strive to minimize potential sources of bias. While the included studies received good quality assessments according to the Newcastle‒Ottawa Scale, the reliance on predominantly observational designs may impact the reliability of the meta-analysis results. Moreover, there was considerable heterogeneity in the included literature, and subgroup and regression analyses were conducted to explore the potential sources of heterogeneity. However, it is important to validate these findings further, as heterogeneity could be influenced by variations in study types. The evaluation of the final level of evidence for the results was not performed, and existing research on trimethylamine N-oxide (TMAO) suggests the presence of sex, age, and ethnic differences. The patients and healthy controls included in this study were predominantly older individuals, while one-third of ischemic strokes occur in young and middle-aged populations. This disparity may introduce changes in the results. Additionally, the majority of the included studies were conducted in China, and the differences between Chinese and Western diets could potentially affect TMAO levels and their association with ischemic stroke. Furthermore, TMAO lacks a standardized reference range or detection method, which may limit the generalizability of the results. Future research should aim to address these limitations by including a more diverse range of study designs, considering different populations and age groups, and standardizing TMAO measurement techniques and reference ranges.Our study offers several notable advantages. Firstly, it stands as the pioneering analysis to incorporate a dose‒response assessment of the association between TMAO and AIS. This unique approach allows for a comprehensive understanding of the relationship between TMAO levels and the risk of AIS, providing valuable theoretical groundwork for future investigations. Secondly, the exclusion of overlapping studies assures that they did not exert a substantial influence on the final results, thereby reinforcing the robustness and reliability of our findings.Our investigation unveiled a compelling association between elevated TMAO levels and the risk of AIS. The dose‒response analysis indicated a significant 12.1% increase in AIS risk for each 1 mol/L increment in peripheral blood TMAO levels. However, it is crucial to exercise caution when generalizing these results due to the considerable heterogeneity and potential biases inherent in the literature. To establish a more comprehensive understanding, further validation through improved research design, larger sample sizes, and inclusion of diverse age groups is warranted. Additionally, meticulous control for confounding factors is essential to elucidate the potential causal relationship between TMAO and AIS.S1 Table(PDF)Click here for additional data file.S2 Table(PDF)Click here for additional data file.S3 Table(PDF)Click here for additional data file.


It has been reported that metformin use may reduce the risk of thyroid cancer, but existing studies have generated inconsistent results. The purpose of this study was to investigate such association between metformin use and the risk of thyroid cancer.I-squared statistic. Combined odds ratios (ORs) with the corresponding 95% confidence intervals (CIs) were calculated through either fixed-effects or random-effects model according to the heterogeneity. Besides, subgroup analyses, sensitivity analyses and test for publication bias were conducted.Studies of metformin use for the risk of thyroid cancer were searched in Web of Science, PubMed, Embase, Cochrane Library, China National Knowledge Infrastructure, China Biomedical Database, Wanfang Data, and Chinese Scientific Journals Database (VIP) from the establishment date to December 2022. Newcastle–Ottawa scale is adopted for assessing the methodological quality of included studies, and the inter-study heterogeneity was assessed by using the P = 0.011). Moreover, in the subgroup analysis, we found that the use of metformin may also aid in the prevention of thyroid cancer in Eastern population  rather than Western population . Sensitivity analysis suggested the results of this meta-analyses were relatively stable. No publication bias was detected.Five studies involving 1,713,528 participants were enrolled in the qualitative and quantitative synthesis. The result of the meta-analyses showed that metformin use was associated with a statistically significant lower risk of thyroid cancer (pooled OR = 0.68, 95% CI = 0.50–0.91, Metformin use is beneficial for reducing the risk of thyroid cancer. For further investigation, more well-designed studies are still needed to elucidate the association between metformin use and the risk of thyroid cancer. Thyroid cancer is a common malignant tumor of the endocrine system, and its incidence is increasing worldwide . AccordiThis study was conducted with the requirements of Preferred Reporting Items for Systematic Review and Meta-Analyses (PRISMA) guideline .A comprehensive search of original studies investigating the association between metformin use and thyroid cancer was carried out in three English databases  and four Chinese databases  to obtain eligible articles from the inceptions of databases to January 2023. We used the following searching terms: (metformin OR dimethylguanylguanidine) AND (thyroid) AND . In addition, the reference lists of the included articles and conference records were also reviewed to identify other potentially inclusive studies.Studies for inclusion must meet the following criteria: (1) the study design should be cohort studies or case–control studies. (2) Metformin use was the intervention or exposure for the exposure group and any feasible intervention without metformin use can be implemented on the control group. (3) The ending indicator was the incidence of thyroid cancer. (4) The study reported relative risks (RRs), hazard ratios (HRs) or odd ratios (ORs) with the corresponding 95% confidence intervals (CIs) or providing sufficient data for calculating the effect size between metformin use and thyroid cancer.Studies with the following characteristics were excluded: (1) the type of articles were not reported or could not be determined. (2) The type of literature was review, in vivo studies, in vitro studies or case reports. (3) The effect size could not be extracted or recalculated. (3) Duplicate studies were also excluded.To ensure the accuracy and objectivity of the data, two evaluators (Hailong Li and Yue Chen) independently screened the literature, extracted information, and evaluated the quality of the studies. These processes need to be cross-checked, and if there was disagreement between the two authors, a third person (Lei Hu) was consulted to assist in the judgment.The articles were screened by first reading the title, and after eliminating obviously irrelevant literature, the abstract and full text were further read to determine the final inclusion. A pre-defined data collection form was used to collected the information. Data extracted from studies included: the first author’s name, publication year, study design, data source, location, sample size, the age of the participants, study period, assessment method of thyroid cancer, variables adjusted in the statistical analyses, quality assessment score, and effect estimates with the corresponding 95% confidence intervals (CIs).The Newcastle–Ottawa Quality Assessment Scale (NOS), a tool for quality assessment of non-randomized studies was employed to evaluate the quality of each study . It consI2 statistic were used to quantitatively assess the extent of heterogeneity of the studies [P < 0.10 and I2 > 50% were considered to be representative of having statistically significant heterogeneity and the random-effects model was then applied for pooling the results. Otherwise, a fixed-effects model was used to create forest plots [The STATA statistical software  was employed to perform the data analyses. Considering the relative low incidence of thyroid cancer, HRs were viewed as ORs, and the results were analyzed through pooling the adjusted ORs with 95% CIs by inverse variance method. Q statistic and  studies . Values st plots . Subgroust plots –31. In ast plots , 33.A total of 480 relevant articles were identified based on the searching terms presented in the method section by initial search across all databases searched. No additional studies were identified through other sources. A total of 154 duplicate articles were removed, and 269 articles were excluded by screening the titles or abstract. During the second stage of full-text screening, we further excluded 52 studies, resulting in five full-text articles , 34, 35 In this meta-analyses, we identified five studies , 34, 35 P = 0.011; Fig. Five articles , 34, 35 P = 0.018) while no such relationship was detected in case–control studies . When stratified by geographic locations, the results were different when confined to different subgroups . For sample size, two studies that followed the study with the sample size ≥ 10,000 indicated a significant impact of metformin use on reducing the risk of developing thyroid cancer . However, in the two studies with sample size of < 10,000, the association between the metformin use and incidence of thyroid cancer was not statistically significant . Further analysis of four high-quality studies regarding its effect on thyroid cancer also find a statistically significant protective effect . There was only one study of moderate quality, which reported no significant reduction in thyroid cancer . All these results of the subgroup analyses are represented in Table In the subgroup analysis by study design, statistically significant associations were observed in cohort studies (pooled OR = 0.61, 95% CI = 0.41–0.92, Z = 0.24, P = 0.806; Egger's test: = 0.48, P = 0.664. Symmetrical funnel plot is shown in Fig. We conducted a sensitivity analysis by changing the random/fixed-effects model, which suggested the statistical stability of this meta-analyses. No publication bias was detected in the analysis of the association between metformin use and dementia/Alzheimer’s disease (Begg's test: Meta-analyses are considered to be an important tool for more accurately assessing the risk of disease development and the effectiveness of treatment , 37. BasSubgroup analysis showed that metformin use was associated with a 45% reduction in thyroid cancer incidence in Eastern countries, compared with only an 11% reduction in Western countries, and this result was not statistically significant in Western countries. Firstly, there is a statistically significant association between metformin and a significant reduction in the incidence of thyroid cancer in Eastern countries compared to Western countries. Thyroid cancer is influenced by a number of factors, including overexpression, mutation or deletion of proto-oncogene sequences , iodine There is a growing body of literature on the beneficial effects of metformin in the prevention and management of cancer Fig. . In addiThe effect of metformin on thyroid cancer has been a hot topic of research in recent years. Metformin inhibits all thyroid cancer cell lines and its effects broadly affect cell proliferation, apoptosis and cell cycle arrest, as well as colony formation and migration . ClinicamTOR pathway targets are activated in thyroid cancer, and metformin may also inhibit the growth, migration and mesenchymal transition of thyroid cancer cell lines through the mTOR pathway other than the insulin pathway . MutatioBoth insulin and IGF-1 can induce cell growth by stimulating the PI3K/AKT signaling pathway . MetformHowever, the meta-analyses we conducted also had certain limitations. Firstly, there were insufficient primary study data included to allow for a more in-depth subgroup analysis of thyroid cancer by gender, age, other health status and specific environmental factors in the populations investigated in both Eastern and Western countries. Therefore, the specific reasons for the analytical finding that metformin use had no significant effect on reducing the prevalence of thyroid cancer in Western countries remain to be investigated, as well as the specific mechanisms by which metformin reduces the occurrence of thyroid cancer in Eastern countries and its exact causes still need to be explored in more depth. Secondly, as all included studies were observational, we were unable to analyze with certainty the specific reasons for the results. Finally, we found a large heterogeneity. However, we performed meta-regression and subgroup analyses to investigate the impact.Despite these limitations, our meta-analyses has some significant advantages. Firstly, the assessment of publication bias showed relatively stable results with insignificant publication bias. Secondly, more studies were included this time than in the previous meta-analyses, which resulted in relatively enhanced statistical power, and we performed a more comprehensive subgroup analysis in this study. In addition, we have provided a detailed account of the mechanisms of metformin action in thyroid cancer, illustrating the possibility of metformin inhibition of thyroid carcinogenesis from a mechanistic review, which also provides a better theoretical basis for the meta-analyses.In conclusion, metformin use may have some effect on reducing thyroid cancer risk. Metformin is a first-line drug for the treatment of T2DM diabetes and also has a therapeutic effect on obesity and ageing-related diseases. Given the high prevalence of diabetes, the yearly increase in the obese population and the increased risk of having thyroid cancer in patients with diabetes as well as obesity, the effect of metformin on thyroid cancer in our study is promising. However, more well-designed basic and clinical studies are needed to further elaborate these associations.


Cancers arising within the gastrointestinal tract are complex disorders involving genetic events that cause the conversion of normal tissue to premalignant lesions and malignancy. Shared genetic features are reported in epithelial-based gastrointestinal cancers which indicate common susceptibility among this group of malignancies. In addition, the contribution of rare variants may constitute parts of genetic susceptibility.A cross-cancer analysis of 38,171 shared rare genetic variants from genome-wide association assays was conducted, which included data from 3,194 cases and 1,455 controls across three cancer sites . The SNP-level association was performed by multivariate logistic regression analyses for single cancer, followed by association analysis for SubSETs (ASSET) to adjust the bias of overlapping controls. Gene-level analyses were conducted by SKAT-O, with multiple comparison adjustments by false discovery rate (FDR). Based on the significant genes indicated by SKATO analysis, pathways analysis was conducted using Gene Ontology (GO), the Kyoto Encyclopedia of Genes and Genomes (KEGG) and Reactome databases.PASSET< 5×10-8). SKAT-O analysis revealed EXOC6, LRP5L and MIR1263/LINC01324 to be significant genes shared by GI cancers . Furthermore, GO pathway analysis identified significant enrichment of synaptic transmission and neuron development pathways shared by all three cancer types.Meta-analysis in three gastrointestinal (GI) cancers identified 13 novel susceptibility loci that reached genome-wide significance (Rare variants and the corresponding genes potentially contribute to shared susceptibility in different GI cancer types. The discovery of these novel variants and genes offers new insights for the carcinogenic mechanisms and missing heritability of GI cancers. AccordiIt is estimated that approximately 5% of GI cancers incidents are attributed to inherited genetic mutations with familial aggregation characteristics , and an Discerning the molecular mechanisms of sporadic GI cancers is complex. In addition to the effort in single cancer research, shared genetic risk factors have been reported in various cancer groups, and target genes with potential cross-cancer roles have been identified at specific genetic loci. For instance, shared mechanisms of carcinogenesis were reported in hormone-sensitive and obesity-related cancers , 13. Com22.1The ESCC, GC, and CRC patients from Fudan University Shanghai Cancer Center (FUSCC) were derived from published studies , which w2.2® Global Screening Array-24 v1.0 Beadchip . Quality control (QC) of the raw genotyping results was conducted in the case-control dataset of each cancer type. The exclusion criteria of SNPs were set as follows: 1) Duplicated variants; 2) Missing rate over 5% between cases and controls; 3) MAF ≥ 1%; 4) SNPs not mapped to autosomal chromosomes; 5) SNPs with genotyping frequency = 0; 6) Deviation from Hardy-Weinberg Equilibrium (HWE) test (P< 1x10-4). The exclusion criteria of ineligible individuals are described as follows: 1) Inconsistent sex between determined genotypes and clinically derived information; 2) First-degree relatives (IBD< 0.5); 3) Missing genotype type > 5%. After QC, 2,158 individuals  and 112,413 variants in ESCC dataset, 2,172 individuals  and 109,771 variants in GC dataset, 2,114 individuals  and 113,867 variants in CRC dataset were retained and subjected to further analysis. The details of the QC process are illustrated in the study flow chart  were found to be shared among the 3 cancer types. ANNOVAR software was used to annotate these rare variants by their location and characteristics . Overall2.4P-values were calculated by additive model adjusted for sex and age. Then meta-analysis of the individual cancer results was conducted to determine the pooled association across these three cancer types, with the significance criteria of genome-wide threshold (P<5×10-8), consistency in risk alleles and odds ratio. The fixed effect model was applied if SNPs passed Cochran’s Q test (Q > 0.1), otherwise random effect models were considered. The association results of individual cancer and meta-analysis were presented by Manhattan plot. Genomic inflation statistics (λ) were calculated and visualized by quantile-quantile (Q-Q) plot to evaluate population stratification in each dataset. To account for the correlation between datasets due to overlapping controls, we applied Association analysis for SubSETs test (ASSET), which is a subset-based approach to determine the best subset of subjects with maximized test statistics for meta-analysis  with 95% confidence interval (CI) and analysis . In our analysis . In orde2.5Due to the fact that the MAF of rare variants in the population is very low, affected individuals may have different mutation sites or frequencies in a particular genetic region of interest. To account for the low efficiency that traditional single variants association analysis may incur, rare variants can be collapsed or compressed into a set and then examined for inter-set frequency differences between case and control groups . A few sb and Qs are the score statistics of the burden test and SKAT test under the null model. The correlation matrix contains a parameter ρ, and mimics the burden test (when ρ= 1) or general SKAT (when ρ= 0). Different genetic structures of different traits correspond to different optimal ρ values. In practical application, ρ is obtained through the test procedure, and the weighted average value of SKAT and burden test statistics is calculated. In this study, the SKAT-O test was applied for gene-based association analysis by using the R package ‘SKAT’ (V2.0.1). For the loci shared by three cancer types, two or more loci located in the same genetic region were combined into a SET (genes)  of less than 0.05 were considered significant.Q (genes) . Additio2.6P-values were adjusted by FDR to control for a low proportion of false positives with the threshold of 0.05. In addition, we calculated Fold Enrichment (FE) by dividing Gene Ratio by Background Ratio to normalize the size of gene set. The shared pathways were identified after pooling the pathway enrichment results of these three GI cancers. Within the shared pathways, we discarded the ‘integral component of postsynaptic density membrane’ and ‘intrinsic component of postsynaptic density membrane’ because they represent protein topology, not a cellular component, and were already obsoleted by the newest vision of the database. For pathways which are subsets of larger pathways, only the larger pathways were retained in our results, the subset pathways were discussed when we explored potential mechanisms of association between shared pathways and GI cancer risk.In order to determine the underlying biological pathways affected by the significant genes identified by SKAT-O analysis, enrichment analysis was performed by Gene Ontology (GO), the Kyoto Encyclopedia of Genes and Genomes (KEGG) and Reactome databases by R package ‘clusterProfiler’ (V4.4.4). For intergenic variants, both annotated genes were included in pathway analysis with subsequent FDR correction. The total gene lists contained 961, 959 and 1,011 genes in ESCC, GC and CRC dataset, respectively. Due to the presence of overlapping genes in common pathways, 2.7P-value of 5x10-8. Manhattan plots and Q-Q plots of each dataset were carried out by R package ‘qqman’ (V0.1.8). Functional annotation of variants was carried out by ANNOVAR based on human genome hg19 coordinates.The comparison of demographic characteristics between the case and control groups was performed by Chi-square test or two independent sample Wilcoxon rank sum test. Variant filtering and quality control were conducted by PLINK (V1.9) and R software (V4.2.0). For single variant association analysis, multivariate logistic analyses were performed by PLINK 1.9 adjusted for sex and age, followed by one-sided ASSET using R package ‘ASSET’ (V2.14.0). Meta-analysis was performed using fixed-effect model followed by Cochran’s Q test. Genome-wide statistical significance was set at 33.1P > 0.05). In the ESCC and GC datasets, over half of the subjects were under 60 years of age, whereas over 50% of the subjects in CRC dataset were over 60 years old. All three datasets showed a higher percentage of male. The distribution of smoking status was significantly different between cases and controls in all three datasets (P< 0.001). More cases in the ESCC dataset were current smokers (57.7%) compared to the control group (36.6%). The majority of the GC and CRC samples were non-smokers, and 38.5% of the patients in the GC’s case group were either smokers or ever-smokers compared to 36.6% in the control group. In the CRC dataset, only 16.2% of the patients were smokers, whereas 32.3% of control samples have smoking history.After quality control of the recruited individuals and genetic data, three case-control datasets of esophageal squamous cell carcinoma (ESCC), gastric cancer (GC) and colorectal cancer (CRC) were included in the study Figure 13.2P< 5x10-4) . In order to assess the shared susceptibility loci across GI cancers, we conducted a meta-analysis to combine the association results of three cancers. Based on one-sided Association analysis based on SubSET (ASSET) analysis (PASSET< 5×10-8), with no pleiotropy or high homogeneity across three cancer types (same direction of effect and Q > 0.1). Of the 13 variants, 8 were likely to be risk alleles (OR >1), while 5 may confer protective effects (OR<1). The most significant variant was LRP5L/rs78345670 (A→G) (P = 9.50×10-5) and CRC dataset . Exonic variant TMEM119/rs112991728 (G→A) .Multivariate logistic regression analysis in ESCC, GC and CRC datasets was performed separately with adjustment for sex and age. A total of 29, 28 and 10 variants were significantly associated with ESCC, GC and CRC risk, respectively (70 (A→G) , which wP<0.05). However, no common variant passed the significant threshold (set as 0.05) in all three validation datasets. Two variants were replicated in two validation populations. EXOC6/rs1339820 – a significant SNP in GC validation , was also significantly associated with CRC in validation . ABCC11/rs75797074 A allele showed protective effect with GI cancer risk in ESCC and CRC validation datasets while the results of ASSET meta-analysis indicated that it was a risk allele for GI cancer initiation .The significant loci in the meta-analysis were further validated in three external datasets  =13.50 (1.77-103.10), P=0.012], whereas older people (>60 years old) may be more susceptible to GC . The opposite effect of age is seen in EXOC6/rs1339820-A holders and risk of ESCC and GC. For holders of the SLIT2/rs147972672-C allele, females have a 2.5-fold increased risk of ESCC compared to males, whereas males are at higher risk for GC. For bearers of RBMS3-AS3/rs73832011-A, the risk for all three GI cancers is only significant in non-smokers. The modifying effect of demographic or lifestyle factors found above is based on the assumption of holding a single variant, for people holding more than one risk variants, this modifying effect may change.The results of stratified analysis showed consistent effects in most subgroups in each cancer type, although the 3.4EXOC6 , LRP5L , MIR1263/LINC01324 were statistically significant in all three cancer types. In addition, two included SNPs EXOC6/rs1339820 and LRP5L/rs78345670 were also significant in the single variant meta-analysis. ABCC11 showed significant in GC and CRC dataset  while the variant rs75797074 located in ABCC11 nearly passed ASSET analysis with the P-value of 5.23×10-8  analysis, 38,171 shared cross-cancer loci were merged into SETs (genes), whereas two or more loci located in the same gene region were merged into a set, and 6,741 sets spanning 28,152 loci were constructed. After FDR correction, 48, 31 and 25 genes were found to be significant in ESCC, GC and CRC, respectively, of which 20 were significant in at least two cancer types Table 3.3.5P-value and most enriched genes in ESCC and CRC datasets. According to the GO database, ‘axonogenesis’ is a subset of the pathway ‘axon development’, in that case, we included the larger pathways in the bubble diagram. Within the CC category, ‘catenin complex’ presented the largest FE in all three cancers , which was also the most significant pathway in CRC dataset.Significant genes in each cancer type identified by SKAT-O were subjected to pathway enrichment analysis based on GO, KEGG and Reactome databases. FDR test was conducted in pathway analysis rather than SKAT-O to avoid over-adjustment. As a result, GO enrichment analysis identified 180, 168, 115 pathways in ESCC, GC and CRC datasets, in which 24 were common across cancers Figure 34Most of the GI cancers share similar multistep carcinogenic processes, therefore exploring shared genetic susceptibility factors can broaden our understanding of the underlying mechanisms in GI cancers. Previous genome-wide association studies focusing on common variants (MAF > 0.01) revealed a handful of loci which associated with individual types of cancer, leaving a considerable proportion of genetic susceptibility remain to be discovered. We hypothesized that parts of the susceptibility may be explained by rare variants with larger effect sizes, and the cross-cancer design may increase the statistical power to detect shared genetic features. This analytical strategy was implemented through meta-analysis along with ASSET method of single variants and supplemented with combined gene-based approach. The biological significance of the identified genes was further explored by pathway enrichment analysis. As a result, we identified 13 novel independent genetic variants associated with GI cancer susceptibility, 20 potential susceptibility genes shared by at least two GI cancers by SKAT-O analysis , and 15 pathways that were significantly enriched within three GI cancers.LRP5L, which is also one of the most significant genes in SKAT-O analysis shared by all three cancers. Another variant rs1339820 located in the 3’-UTR region of EXOC6 which was also a member of shared associated genes in GI cancers. This variant was also significant in GC and CRC validation datasets, although only the OR orientation in GC was consistent with the original meta-analysis. ABCC11/rs75797074, a significant SNP associated with risk of all three cancers (yet not passed the ASSET testing threshold), was successfully validated in the CRC and ESCC datasets  is a key gene involved in the regulation of Wnt signaling pathway, which plays an important role in tumor progression  tissues and associated with poor survival in OV patients. Overexpression of this gene also promoted proliferation, invasion, and migration in OV cells  – one of the susceptibility genes indicated by single variant analysis, acts as a member of glutamatergic synaptic transmission pathway which may contribute to GI cancers carcinogenesis by regulating glutamate receptor in synapse (LRRN4 and NLGN1 in the nerve or glutamatergic synapses complex were found to promote CRC progression. The transmembrane protein Neuroligin 1 (NLGN1) is reported to interact at the synapse with the tumor suppressor Adenomatous Polyposis Coli (APC), which is intensively involved in the pathogenesis of CRC and is a key player in the WNT/β-catenin pathway  modulate part of GI tract selectively , among w synapse . Based o pathway . As for  pathway –50. Gene pathway . The syn pathway . Former er cells . Their rSLIT2 might mediate the process of axon guidance by attracting or repelling developing axons and migrating neurons (Neuron density is generally regarded higher in tumor tissues than normal tissues , and neu neurons . In our  neurons . Moreove neurons . To thisThere are several limitations to our study. First, due to the retrospective nature of our patient cohort, missing information may affect the accuracy of our multivariate logistic regression analysis, which only adjusted for age and sex. Important lifestyle factors such as drinking cannot be assessed. Second, the sample size of each cancer type was relatively small, with the overlapping control population. Although we tried to address this potential issue by using the ASSET test, this may still result in limited statistical power. However, the research interest of cross-cancer susceptibility has allowed us to focus only on the shared association through the meta-analysis. Last, our study population come from the eastern China, while the validation dataset included subjects from the northern and central China. No single SNP was validated in all three validation datasets suggests potential population stratification hence extrapolation of the result needs cautious implementation.5Based on a sequential investigation of single variant, gene-based and pathways enrichment analysis, we uncovered novel rare variants and genes that may contribute to the susceptibility of GI cancers. Further studies are warranted to look into the underlying mechanism of the association between these susceptibility markers and GI cancers.The summary data of the detailed results of the study are included in the article/All patients provided an informed consent to participate in the study, and the study protocol was approved by the ethical committee of the Institutional Review Board of Fudan University Shanghai Cancer Center.The hypothesis of analyzing shared rare variants associated with the risk of GI cancer was conceived by RZ. JZ conducted quality control and bioinformatic analysis. Data collection and the bioinformatic analysis was conducted by XW. JL helped with the visualization of results. QW provided suitable guidance to the study design. MYW and YW helped with the collection of blood DNA samples. JC, JX, MLW and TW assisted with validation of significant SNPs in external datasets. All authors have read and approved the final manuscript.