Rotavirus vaccination has been proven to effectively protect against rotavirus gastroenteritis. However, there are concerns about the relationship between rotavirus vaccination and the risk of autoimmune disorders. Thus, we conducted a systematic review and meta-analysis to comprehensively assess the association between rotavirus vaccination and type 1 diabetes (T1D) or celiac disease (CD) risk.A systematic review and meta-analysis were conducted to evaluate the type 1 diabetes or celiac disease associated with rotavirus vaccination. The following journal databases were searched to identify potential studies for inclusion: PubMed, Embase, and Cochrane Library databases.Seven articles involving more than 5,793,055 children were included. Our results showed that rotavirus vaccination does not alter the subsequent risk of T1D  or CD  after vaccination. Furthermore, the risk of T1D was not increased or decreased for children fully exposed to rotavirus vaccination  and for children partially exposed to rotavirus vaccination . However, younger (<5 years) vaccinated children at the end of study  may be at a lower risk for T1D than older (≥5 years) vaccinated children .The findings of this study suggest that rotavirus vaccination does not appear to be associated with T1D or CD in children. The protective effect of rotavirus vaccination on T1D may be presented by time dependent. Rotavirus infection is the most common cause of acute gastroenteritis in children under 5 years of age worldwide . In 2009Rotavirus infection is a risk factor for type 1 diabetes (T1D)  and celiThis systematic review and meta-analysis are reported in accordance with the Meta-Analysis for Observational Studies in Epidemiology (MOOSE) checklist  Supplem and PrefObservational studies were included if they were published as a peer-reviewed article; had a cross-sectional, case-control, or cohort design; compared the risk of T1D or CD between rotavirus vaccine exposure and non-exposure groups; reported incidence rate ratios (IRRs), odds ratios (ORs), relative risks (RRs), or hazard ratios (HRs); and provided adequate data to allow calculation of risk estimates when adjusted data were not provided. Case reports, case series, animal studies, editorials, and reviews were excluded.Data were independently extracted by two authors; any discrepancies were resolved through discussion with another author. The extracted data included the first author’s name, publication year, study design, study location, study period, participant characteristics, method of diagnosis of T1D or CD, statistical adjustments, and study quality. In cases where more than one estimate was provided, the most adjusted effect size estimates reported by each study were used. We assessed the methodologic quality of the included observational studies based on the Newcastle-Ottawa Scale (NOS) as recommended by the Cochrane Collaboration . The sca2 test and I2 statistic; an I2 of >50% or p-value < 0.05 for the Q-statistic was considered to indicate substantial heterogeneity . Statistical heterogeneity of the included studies was calculated by the χogeneity . Data weogeneity . Becauseogeneity . Publicaogeneity . A p-valAfter excluding duplicate reports, we identified 166 citations. Based on the titles and abstracts of the papers, 143 were excluded, and the remaining 23 were evaluated on the basis of the full-text articles. Seven studies met the inclusion criteria. Some of the excluded studies, along with reasons for their exclusion, are shown in p = 0.41; 2 = 43.7%). Analysis of the cohort studies alone indicated that the combined RR of T1D was 0.95 . The RR of T1D was 0.86 . In analyses by age at the end of study, vaccinated children aged less than 5 years were at a decreased risk for T1D  and 0.86 (95% CI: 0.64–1.17), respectively. However, a decreased risk of T1D was observed in vaccinated children aged less than 5 years at the end of study.Rotavirus is the most common viral cause of diarrhea in children . A numbeT1D and CD have high heritability, estimated to be 70 and 80%, respectively, from twin studies , 30. TheAn unanticipated finding was that vaccination in children aged under 5 years was associated with a decreased risk for T1D, implying that rotavirus vaccination has a time-dependent effect on T1D risk. This may result for two reasons. First, rotavirus vaccine efficacy was proved to wane with time in a recent meta-regression study . TherefoThis systematic review and meta-analysis is the first to provide an overall estimate of the effects of rotavirus vaccination on the subsequent risk of T1D and CD. Our meta-analysis has the advantage of the exclusive inclusion of cohort studies, which are less prone to bias. However, this study had several major limitations. First, there are an unknown number of residual confounders. Further well-designed studies that consider more covariates, such as family history of T1D or CD and breastfeeding, are required to evaluate the association between rotavirus vaccination and the risk of T1D and CD. Second, the duration of follow-up in the included studies was not long enough to determine possible long-term effects. Third, the number of eligible studies and sample size for CD were small, which may have influenced the accuracy of our results. Also, no study reported adjusted estimates by gender, which is an important distinction, as boys and girls are known to have differential risks of CD . Future In conclusion, the results of this meta-analysis failed to demonstrate a role for rotavirus vaccination in the development of CD in children. However, we could not rule out that the protective role of rotavirus vaccination in T1D development was time dependent and further studies are still needed to verify our findings. Our findings support continued worldwide rotavirus vaccination, reducing the burden of rotavirus morbidity in those populations. As there was a small number of studies included in our review, continued evaluation of this association is warranted, and there is a pressing need for new studies with longer follow-up to further explore the relationships between rotavirus vaccination and the risk of T1D and CD in older children.The original contributions presented in this study are included in the article/XZ and JJ conceived the study and revised the manuscript critically for important intellectual content and made substantial contributions to the design, acquisition, analysis and interpretation of data. X-FX participated in the design, selection, and analysis and interpretation of data. All authors read and approved the final manuscript.


Background: Numerous studies have demonstrated an association between osteoprotegerin (OPG) polymorphisms (A163G (rs3102735), T245G (rs3134069), T950C (rs2073617), G1181C (rs2073618)) and osteoporosis risk. However, their conclusions are inconsistent. In addition, some new studies have been updated, and more importantly, previous meta-analyses have not tested for false-positive results. In order to further explore these associations, we recently conducted a meta-analysis.Objectives: To study the relationship between OPG polymorphisms A163G, T245G, T950C, G1181C and the risk of osteoporosis.Methods: PubMed, Medline, International Statistical Institute (ISI), China National Knowledge Infrastructure (CNKI) and China Wanfang Database were used for research searches. Associations were assessed with five genetic models using odds ratios (ORs) with 95% confidence intervals (CIs). In addition, confidence in statistically significant associations was assessed using false-positive report probability (FPRP), Bayesian probability of False discovery (BFDP), and Venice criteria.Results: On the whole, the OPG A163G polymorphism was not significantly associated with risk of osteoporosis. However, in a subgroup analysis, we found that the OPG A163G polymorphism increased the risk of osteoporosis in Caucasians  and the female . At the same time, the OPG G1181C polymorphism reduces the risk of osteoporosis . Moreover, a significantly decreased risk of osteoporosis was also discovered in Asian  and the female . Finally, we did not find a close association between OPG T245G and T950C polymorphisms and osteoporosis risk. However, when we retained only studies in the control group that was consistent with Hardy-Weinberg equilibrium (HWE) and high-quality scores, we observed that the OPG A163G polymorphism increased the risk of osteoporosis in the overall analysis . Finally, after the credibility assessment, we concluded that all statistically significant association results in the meta-analysis in this study and those in the previous study were ‘positive results with low confidence’.Conclusion: In conclusion, our study concluded that all meaningful results between OPG A163G and G1181C polymorphisms and osteoporosis risk were false-positive results rather than true associations. Osteoporosis is a systemic bone disease characterized by deterioration of femoral microstructure and low bone mass, which increases fracture susceptibility and bone fragility . The WorITGA1, ESR1, SPP1, LRP4, and LRP5 have been confirmed to be related to bone matrix composition, bone mineral density (BMD) homeostasis and bone remodeling, thereby affecting BMD , but the association of these polymorphisms with osteoporosis risk has not been thoroughly demonstrated , also called osteoclast genesis inhibitory factor, is a secreted glycoprotein discovered by Simonet et al., in 1997. It is a single copy gene cluster located on human chromosome 8q24.2. So far, many researches have shown that OPG gene has relationship to the pathogenesis of osteoporosis , T950C (rs2073617), and T245G (rs3134069), which are located at the promoter of the gene, while G1181C (rs2073618) is located at the first exon of the gene. Although researchers have published meta-analyses of OPG and osteoporosis risk. . The literature search was conducted until 20 March 2022. In addition, we carefully reviewed each reference list for meta-analyses in the literature to identify all eligible studies.The literature was searched using PubMed, Medline, International Statistical Institute (ISI), China National Knowledge Infrastructure (CNKI) and China Wanfang Database. The retrieval strategy was as follows  AND (osteoporosis OR osteoporoses) AND  to describe the relationship between We designed the data extraction form in advance. All literatures were screened according to the inclusion criteria and exclusion criteria, and then the literature data were extracted for statistical analysis. When the two authors obtained different results and still could not reach a consensus after discussion, the third author extracted the data again, and finally the three authors jointly checked and confirmed. If the data in the article is unclear or disputed, we will contact the original author to obtain the original data. The extracted information included: year of publication, first author of the article, country of study, ethnicity of study subjects, sex of study subjects, female menopause, polymorphism genotyping technique, type of control group, sample size and mean age for case and control groups, and quality score of all articles .The quality assessment for all included articles was assessed separately by two authors. We referenced and improved the specific criteria for quality scores from previous studies . The fulOPG gene polymorphisms  and osteoporosis was expressed by ORs and corresponding 95% confidence intervals. The following five genetic models were selected for evaluation: 1) allele model ; 2) dominant mode (homozygous variant + heterozygous variant vs homozygous wildtype); 3) recessive model (homozygous variant vs heterozygous variant + homozygous wild type); 4) over-dominant model (homozygous wild type + homozygous variant vs heterozygous variant); 5) additive model (homozygous variant vs homozygous wild type). Chi-square q test and I2 value were used to evaluate the heterogeneity of the results. When p was less than 0.10 and/or I2 was greater than 50% , nine studies investigated OPG T245G , 12 studies investigated OPG T950C , and 18 studies reported OPG G1181C , as shown in OPG A163G had three low-quality studies and 11 high-quality articles, OPG T245G had two low-quality articles and seven high-quality articles, OPG T950C had five low-quality articles and seven high-quality articles, and OPG G1181C had two low-quality articles and 16 high-quality articles. The complete characteristics and genotype frequencies of the final included literatures are shown in At the beginning of the period, 369 articles were searched from ISI, PubMed, Medline, CNKI and the China Wan-fang databases. There are 152 records left after duplicates are removed. After careful screening for titles and abstracts, we excluded 87 papers. In addition, 17 items were excluded because of data duplication or the inability to extract detailed data, and 18 items were excluded because of inadequate controls. Finally, 30 articles and 31 studies were included in our study . A totalOPG A163G polymorphism and the risk of osteoporosis. However, in the following ethnic subgroup analysis, we found that the OPG A163G polymorphism increased the risk of osteoporosis in Caucasians  and HWE (G vs A: p = 0.010) were the source of heterogeneity between the OPG A163G polymorphism and the risk of osteoporosis. The type of controls (TG + GG vs. TT: p = 0.006) was the sources of heterogeneity between the OPG T245G polymorphism and the risk of osteoporosis. At the same time, the female menopause , sample size (TC + CC vs. TT: p = 0.037) and sex (TT + CC vs. TC: p = 0.045) were the sources of heterogeneity between the OPG T950C polymorphism and the risk of osteoporosis. For the OPG G1181C, no covariate was found as a possible cause of between-study variation.During the statistical process, we found several possible sources of heterogeneity, including ethnicity, sex, sample size, female menopause, control type, quality score and HWE. Therefore, we used meta-regression analysis to determine the causes of heterogeneity. A meta-regression analysis revealed that the sample size  FPRP <0.2 and BFDP <0.8. If all four criteria are not met, then all meaningful results are considered “less credible”. Our study also used BFDP, FPRP and Venice criteria to evaluate the statistically significant credibility. Associations that met the following criteria were considered to have high confidence: 1) at least two genetic models were statistically significant; 2) statistical power >80%; 3) OPG polymorphism has been considered one of the potential genetic factors for osteoporosis. The OPG/RANKL/RANK is the main signal transduction pathway in osteoporosis, which can regulate the differentiation, induction, activation and maintenance of osteoclasts , in other words, all statistically significant results were considered false-positive results.Overall, we did not find a significant association between parisons . TherefoOPG G1181C polymorphism and risk of osteoporosis. As far as we know, some low quality and small sample sizes studies are common bias and errors. These studies are not always strict and the quality is poor, so the statistically significant small sample study is more likely to be accepted, it will be very easy to produce false positive results, making the conclusions of these original studies implausible. In the test of publication bias, the asymmetry of the funnel plot is caused by the study of low quality and small samples.Significant publication bias was only observed between the OPG A163G, T245G, T950C and G1181C polymorphisms and osteoporosis. There is an obvious mismatch between all previous meta-analyses and this meta-analysis in the subgroup analysis classification. Moreover, after careful reading of the full text, we found that the previous meta-analysis had relevant information and data extraction errors in the process of data extraction, and we corrected them in the extraction process. In addition, the sample size of our study is large. A total of 31 studies were applied in this study , of which 14 studies showed the OPG A163G , nine studies investigated the OPG T245G , 12 studies investigated the OPG T950C , and 18 studies reported the OPG G1181C . In this study, five genetic models were compared separately. Li et al., 2017, however, applied four genetic models; Li et al., 2021, used only three genetic models. At the same time, Li et al., 2017, the ethnicity studied was limited to Chinese. It is important to note that previous meta-analyses did not perform credibility analyses on the results, and all statistically significant relationships were considered “less credible” when we assessed the credibility of previous meta-analyses using the same criteria. Therefore, their research results and conclusions may not be reliable, and they may also be false positive results.All four previous meta-analyses researched the correlation between Compared with previous meta-analyses, our study has the following advantages: 1) FPRP, BFDP test and Venice criterion were used for credibility assessment; 2) to evaluate the quality of qualified research; 3) compared with previous meta-analyses, the studies included in this study are more comprehensive, and the data collected are more detailed and accurate; 4) the subgroup analysis was more comprehensive, and it is important to note that not only did we subgroup sex, but we also grouped women separately based on whether they were postmenopausal or not; However, our study still has some potential limitations. First, the current meta-analysis included only published studies, and it is well known that positive results are studied more than negative results. Second, osteoporosis is a complicated polygenic disease, and individual SNPS are slightly associated with osteoporosis risk. However, we have not retrieved data on gene-gene and gene-environment combined effects. Third, although we objectively collected data from the included literature, the control: control rate was less than 1, which may also affect the accuracy of the association. At the same time, there have been relatively few studies of Africans in ethnic analyses and insufficient statistical power to excavate true associations. Therefore, future studies with large samples and sufficiently large subgroups will help validate our findings. Finally, the risk of osteoporosis includes not only genetic factors, but also environmental and lifestyle factors. Our study did not control for various osteoporosis variables, such as smoking, alcohol consumption, breastfeeding, physical activity, estrogen replacement therapy, corticosteroids, calcium supplements et al. In the future, we need to control for various other variables that contribute to osteoporosis, and investigate the association between osteoporosis and genetic, environmental, and lifestyle factors to help validate our findings.OPG A163G and G1181C polymorphisms and osteoporosis risk were false-positive results rather than true associations. At present, there is still no absolute strong evidence for the real association between OPG polymorphism and osteoporosis risk, and further large-scale epidemiological studies are needed to confirm or deny our findings.In conclusion, this study indicates that all meaningful results between


To present an update of the available literature on external beam radiation therapy (EBRT) with or without brachytherapy (BT) compared to radical prostatectomy (RP) for patients with high-risk localised prostate cancer (PCa).2I statistic, respectively. We performed six meta-analyses for overall survival (OS) and cancer-specific survival (CSS).We conducted a systematic review and meta-analysis of the literature assessing the survival outcomes in patients with high-risk PCa who received EBRT with or without BT compared to RP as the first-line therapy with curative intent. We queried PubMed and Web of Science database in January 2021. Moreover, we used random or fixed-effects meta-analytical models in the presence or absence of heterogeneity per the A total of 27 studies were selected with 23 studies being eligible for both OS and CSS. EBRT alone had a significantly worse OS and CSS compared to RP . However, there was no difference in OS  and CSS  between EBRT plus BT compared to RP.While cancer control affected by EBRT alone seems inferior to RP in patients with high-risk PCa, BT additive to EBRT was not different from RP. These data support the need for BT in addition to EBRT as part of multimodal RT for high-risk PCa.Abbreviations: ADT: androgen-deprivation therapy; BT: brachytherapy; CSS: cancer-specific survival; HR: hazard ratio; MFS, metastatic-free survival; MOOSE: Meta-analyses of Observational Studies in Epidemiology; OR: odds ratio; OS: overall survival; PCa: prostate cancer; RR: relative risk; RP: radical prostatectomy; RCT: randomised controlled trials; (EB)RT:  radiation therapy High-risk non-metastatic disease  accounts for ~30% of newly diagnosed prostate cancer (PCa) ,2, depenIn the current and likely future, the absence of propensity designed RCTs comparing RP to RT for patients with high-risk PCa, meta-analyses may help a framework for clinical decision-making and patient counselling. To this end, we performed a systematic review and meta-analysis to summarise the results of available studies including the latest literature on this subject. We focussed on the additive value of BT in addition to EBRT.In this meta-analysis, we followed the Meta-analyses of Observational Studies in Epidemiology (MOOSE) statement guidelines that propose a checklist of items which provides a RCTs checklist . MoreoveThe question of this study was, ‘Do patients with high-risk PCa who receive EBRT with or without BT have a better survival outcome compared to those who receive RP?’. All current articles covering the study question were eligible for this systematic review. We selected studies that perform quantitative synthesis according to the similarity in Population, Interest, Context (PICO) elements to decrease the selection bias and heterogeneity. The inclusion criteria for the quantitative meta-analysis were original research articles that assessed survival outcomes and reported an estimated risk effect  for both patient and control groups. Exclusion criteria were BT usage alone as definitive therapy and lack of definition of high-risk PCa in the main or subgroup analysis.Consequently, the more comparable cohort studies according to the MOOSE guidelines were included in the analyses. Furthermore, the heterogeneity of the population was explored by detecting the source and country of databases. According to OS, we categorised studies’ outcomes, CSS, biochemical recurrence survival, and metastatic-free survival (MFS).We searched PubMed and the Web of Science for studies published before 1 January 2021. The search queries line and search strategies were ‘((Prostate OR prostatic) AND (cancer OR carcinoma) OR (Prostatic Neoplasms[Mesh])) AND external beam radiotherapy OR brachytherapy OR radiotherapy [Mesh]) AND ’ in PubMed and ‘((Prostate OR prostatic) AND (cancer OR carcinoma)) AND  AND ’ in the Web of Science.http://cccrg.cochrane.org/author-resources). We extracted the following data: first-author, type of article, year of publication, dates of the data collection or enrolment, cohort type, sample size, number of individuals on treatment, outcome, how the outcome was measured, type of effect statistic, effect statistic error measures, and effect statistic P value. There were no limitations in the articles’ data, so we did not need to contact any authors for additional details. Modified Newcastle-Ottawa Scale criteria were used to assess the quality of the included studies [The search results were restricted to English language articles. Two reviewers screened titles and abstracts independently; any disagreement about the articles’ eligibility was resolved by Delphi consensus with the co-authors. A data extraction sheet was developed based on the Cochrane Consumers and the Communication Review Group’s . Heterogeneity across the studies was appraised using P values, Q and 2I statistics [P value was <0.05. All analyses were carried out using Stata version 14 .Forest plots were used to assess the multivariable HRs. We summarised them to depict the relationship of our outcomes with the type of treatment . When HRs and atistics . In the After initial screening, 526 articles were available for assessment. The selection process for the systematic review is shown in Almost all of the studies in this review were cohort studies. Of the 27 included studies, regardless of the type of RT , 23 studies assessed OS and CSS. The quality assessment of the included studies according to the Newcastle-Ottawa scale is summarised in I2 = 84.5%, P < 0.001), so a random-effect model was used. The funnel plot was asymmetrical . In the second meta-analysis, 23 studies were again included. We found that patients with high-risk PCa who received RT (regardless of the type of RT) had significantly worse CSS than those treated with RP with a HR of 1.37 , so a random-effect model was used. The funnel plot was asymmetrical .In the first meta-analysis of OS, 23 were included. We found that patients with high-risk PCa who received RT (regardless of the type of RT) had significantly worse OS than those treated with RP with a HR of 1.27 , so a random-effect model was used. The funnel plot was slightly asymmetrical . In the second subgroup meta-analysis, 18 studies were included. We found that patients with high-risk PCa who received EBRT alone had significantly worse CSS than those treated with RP with a HR of 1.55 , so a random-effect model was used. The funnel plot was slightly asymmetrical .In the first subgroup meta-analysis, 15 studies were included. We found that patients with high-risk PCa who received EBRT alone (i.e. without a concomitant BT) had significantly worse OS than those treated with RP with a HR of 1.38 , so a random-effect model was used. The funnel plot was asymmetrical . In the fourth subgroup meta-analysis, seven studies were included. The HR of 0.69 (95% CI 0.45–1.06) suggested no difference in CSS between patients with high-risk PCa who received EBRT plus BT and those who underwent RP; however, statistical significance was not reached , so a random-effect model was used. The funnel plot was slightly asymmetrical . In the third subgroup meta-analysis, eight studies were included. The HR of 1.1 (95% CI 0.76–1.34) suggested that there was no difference in OS between patients with high-risk PCa who received EBRT plus BT compared to those who underwent RP . The eig reached . The sevThe present systematic review and meta-analyses assessed the comparative survival effectiveness of RT and RP as a definitive therapy with curative intent of PCa patients with high-risk features. Although we found that RT irrespective of RT type  resulted in significantly worse OS and CSS compared to RP, patients who underwent a EBRT plus BT combination had OS and CSS that were not inferior to RP. However, EBRT alone (i.e. without concomitant BT) was inferior to RP with regards to OS and CSS.While there is no direct, well designed comparison of RP vs RT as the first step in a multimodal therapeutic concept in concordance with our meta-analyses, cumulative data support the concept of combining maximal RT consisting of EBRT plus BT with androgen-deprivation therapy (ADT) as the first-line multimodal strategy for therapy of high-risk PCa ,17,19,29Today, RT as adjuvant or salvage strategy is supported as a part of a multimodal therapy after RP by major guidelines for patients with high-risk PCa based on data recruiting from several RCTs . HoweverThe main limitation of the present systematic review and meta-analysis was the lack of well-designed controlled trials. However, because of this lack, we believe that this systematic review and meta-analysis might help frame and equipoise decisions to guide patients’ counselling as part of the shared decision process . AnotherAccording to this systematic review and meta-analyses, patients with high-risk PCa who received EBRT alone (without a concomitant BT) as a first-line definitive therapy had worse OS and CSS than those who underwent RP. However, EBRT plus BT as a multimodal RT was not inferior to RP in high-risk PCa. These data support the need for a multimodal strategy to achieve optimal therapy in high-risk PCa. RT as a primary definitive strategy could be better when combined with EBRT and BT to achieve maximal radiation dose combined with ADT. The role, timing, indication of postoperative RT after RP, and the type and duration of ADT for each patient need assessment and evidence. Until then, we would postulate based on the findings of our present study an equipoise of EBRT with BT for local control as a part of multimodal flexible and dynamic treatment strategy tailored to each tumour in each patient.Click here for additional data file.


The prognostic value of the HS-mGPS for the OS trended in a consistent direction after subgrouping and sensitivity analysis. In conclusion, the HS-mGPS serves as a valid prognostic biomarker for cancer patients, with a high HS-mGPS associated with adverse survival outcomes.The suitability of the high-sensitivity modified Glasgow Prognostic Score (HS-mGPS) in cancer patients remains unknown. We performed a systematic database search from 1 January 2010 to 30 September 2022, in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses guidelines. Selected studies reported the HS-mGPS and survival outcomes in cancer patients. The association between the HS-mGPS and survival outcomes was evaluated using a random-effects model and expressed as pooled hazard ratios (HRs) with 95% CIs. This meta-analysis evaluated 17 studies with a total of 5828 cancer patients. A higher HS-mGPS was found to be associated with an adverse OS , DSS , and DFS (HR = 1.96; 95% CI: 1.48–2.58; all  Cancer constitutes one of the leading causes of mortality worldwide and is responsible for nearly 10 million deaths per year; one in five people develop cancer during their lifetime, with an estimated 19.3 million new cases detected in 2020 . In addiRoxburgh and McMillan validated the role of systemic inflammatory response in the survival predictions for patients with primary operable cancer in 2010 . PreviouClinicalTrials.gov to identify relevant ongoing trials. Institutional Review Board approval was waived because no individual data were used.This systematic review and meta-analysis followed the Preferred Reporting Items for Systematic Reviews and Meta-Analyses guidelines . This meThe study inclusion criteria were as follows: (1) enrolling patients with all cancer types; (2) investigating the association of the HS-mGPS with survival outcomes; and (3) providing hazard ratios (HRs) with 95% CIs or sufficient information to enable their calculation. The following were excluded: (1) reviews, meta-analyses, conference abstracts, case reports, letters, or commentaries; (2) studies lacking data on survival outcomes; and (3) studies enrolling patients who had participated in a previously published trial. Two authors (Luan CW and Wu TH) independently screened all study titles and abstracts, excluded irrelevant studies, and reviewed the full texts of the remaining studies. A third author (Yap WK) was consulted in cases when the aforementioned authors could not reach a consensus.Two independent authors (Luan CW and Wu TH) extracted the following data from the evaluated studies: (1) study characteristics ; (2) participant characteristics ; (3) selected cutoff values of the HS-mGPS for survival analysis; and (4) HRs with 95% CIs for overall survival (OS), disease-free survival (DFS), and disease-specific survival (DSS). If a study provided the HRs with 95% CIs of an HS-mGPS of 1 versus 0 and an HS-mGPS of 2 versus 0 as two groups, we calculated the data as two datasets. If a study presented the results in both multivariable and univariable analysis, the adjusted HRs of the multivariable analysis were used for analysis . Formal We used the Newcastle–Ottawa Quality Assessment Scale (NOS), which consists of eight main elements for study quality evaluation , when asp value of <0.05. We used pooled HRs with 95% CIs to quantify the primary study outcomes , and the results of the pooled data were adopted to investigate the associations between the HS-mGPS and patient prognoses. An I2 test and Cochran’s Q test were also conducted to assess heterogeneity between studies. In Cochran’s Q test, statistical significance was indicated at p < 0.1. I2 values of ≤24.9%, 25% to 49.9%, 50% to 74%, and ≥75% indicated no, low, moderate, and high heterogeneity, respectively [A random-effects model was applied to incorporate survival outcomes in the current meta-analysis due to the expectation of heterogeneity within the included studies . Comprehectively . We perfectively .The literature search and selection process are summarized in The general characteristics of the included studies are summarized in p < 0.001) with moderate heterogeneity  and soft tissue sarcoma (I2 = 84.0%) but not in head and neck cancer (I2 = 0%) suggested that the prognostic value of the HS-mGPS may vary by tumor type. We observed a significant and constant association between the HS-mGPS and OS in the subgroups based on research region, sample size, HS-mGPS cutoff value, and analysis method; these results support the consistency of our findings and indicate that a high HS-mGPS likely predicts poor OS. Mild-to-moderate heterogeneity was detected across the included studies (I2 = 36.15–78.90%), with the exception of the cutoff value = 1 subgroup (I2 = 0%). The varied heterogeneity across subgroups may provide additional insight into the possible causes of heterogeneity observed in the pooled results.Subgroup analyses were conducted to further examine the consistency of the prognostic role of the HS-mGPS for OS. We divided the 20 datasets groups based on tumor site, research region, sample size, HS-mGPS cutoff value, and analysis method. The results are presented in p < 0.001) with no heterogeneity . No significant publication bias was detected by Begg’s test and Egger’s test. Subgroup and sensitivity analyses were not performed because of the limited number of available studies.Three studies ,31,33 wip < 0.001) and Egger’s test (p < 0.001). Therefore, we used the trim-and-fill method to evaluate the influence of publication bias on the pooled results. The results from the trim-and-fill calculation indicated a pooled HR of 1.63  for OS prediction after adding the missing studies. The smaller adjusted HR did not alter the direction or significance of the results, a finding that supports the validity of this meta-analysis. In addition, sensitivity analyses were conducted through the individual omission of studies from the pooled analysis to assess the stability of the results with regard to OS , suggesting an association between high HS-mGPS and poor OS outcome. Similarly, a high HS-mGPS in cancer patients served as an unfavorable prognostic factor for DFS and DSS. This study is the first systematic review or meta-analysis to demonstrate the prognostic role of the HS-mGPS with respect to survival outcomes.This meta-analysis combined data from 15 studies (5545 patients) and determined a statistically significant association between the HS-mGPS and OS of cancer patients; statistical significance was maintained in sensitivity analyses. The background conditions of patients varied but did not exert a statistically significant effect on the direction of the associational between the HS-mGPS and OS. The subgroup analysis results indicated that the prognostic value of the HS-mGPS for OS maintained significance for cancer type, sample size, cutoff value of the HS-mGPS, research region, and analysis method . As we kGrowing evidence has revealed that ratios based on components of the white cell count are positively associated with tumor malignancy, including the neutrophil-to-lymphocyte ratio, the platelet-to-lymphocyte ratio, the monocyte-to-lymphocyte ratio, and the Systemic Immune–Inflammation Index ,35. FurtHow the HS-mGPS is relevant to adverse prognoses in cancer patients remains undetermined. An elevated HS-mGPS is indicative of a patient with hypoalbuminemia, a high CRP level, or both. CRP acts as an acute phase plasma protein in response to inflammation or infection; it is regulated by proinflammatory cytokine stimulation such as tumor necrosis factor-alpha, interleukin-1 (IL-1), and IL-6 ,40. In aThe staging system, an essential element of the cancer treatment strategy, nevertheless excludes individual factors. Clinicians explain to patients that individual factors account for the varying prognoses of patients who are at the same stage but with different prognoses. With this fact in mind, a simple, readily available, and reliable score system that can generate valuable prognostic data may serve as helpful treatment information for clinicians. The development of personalized treatment regimens is of vital importance; therefore, many studies have discovered better prognosis tools by focusing on a combination of a staging system with the immunological and nutritional status of the patient. The results of this meta-analysis summarized the existing evidence of the effectiveness of the HS-mGPS in survival prognosis and endorsed its prognostic value for cancer patients. However, several studies that we analyzed which compared the mGPS and the HS-mGPS indicated a higher prognostic value for the HS-mGPS ,15,16. AThis meta-analysis has several limitations. First, all included studies were retrospective and displayed considerable methodological diversity. The inclusion of various cancer types, cancer stages, and treatment strategies may have contributed to the high statistical heterogeneity observed in our meta-analysis. However, the sensitivity test and subgroup analysis displayed no statistically significant effect on the direction of the association between the HS-mGPS and survival outcomes. Second, there were no published reports from American or European countries, with all included studies examining Asian populations. The global application of our findings remains undetermined. Moreover, the meta-analysis had statistically significant publication bias; however, a trim-and-fill analysis did not alter the direction or significance of the results. Finally, the HRs obtained in the studies were calculated with a combination of multivariable analysis (13 studies) and univariable analysis (4 studies), which may have influenced the reliability of the findings. Therefore, we advise others to interpret and use our results cautiously, and we encourage further prospective studies to reinforce our findings.The current study suggested that high HS-mGPS has an adverse effect on survival in cancer patients. This meta-analysis reviewed the existing evidence of the usefulness of the HS-mGPS as a biomarker and confirmed its reliability for predicting cancer prognosis. Future large-scale, prospective clinical trials are required to further validate our study results in the global population.


Many laboratory studies have shown that tea consumption protected against the development of esophageal cancer (EC). However, in epidemiological studies, inconsistent or even contradictory results were frequently observed, especially when drinking tea at higher temperatures.We conducted a meta-analysis based on published observational studies to explore whether hot tea consumption was a risk factor of EC. Relevant studies were searched in PubMed, Embase, and Web of science up to October 13, 2021, and we also manually retrieved the literature in the included studies and recent reviews.p = 0.00). Except for esophageal adenocarcinoma (EAC), this increased risk was also found in the majority of subgroups, which are the European and Australian populations.A total of 23 eligible reports were identified, including 5,050 cases and 10,609 controls, and a meta-analysis with Comprehensive Meta-Analysis (CMA) software (version 2.0) was conducted. A statistically significant increased EC risk was observed when drinking tea at higher temperature (odds ratios (ORs) = 1.79, 95% CI: 1.48–2.15, This meta-analysis showed that people who drank hot tea had a significantly increased risk of Esophageal squamous cell carcinoma (ESCC), but no significant association for EAC. Esophageal cancer (EC) is the seventh most commonly diagnosed cancer and the sixth cause of cancer mortality worldwide, with approximately 70% of cases occurring in men and a 2- to 3-fold difference in the incidence and mortality rates between different regions . EsophagAlthough the definitive mechanism of EC development is still unclear, many studies have shown that dietary habits are significantly associated with the development of EC –9. For eTea beverages are usually a mixture of boiling water and tea leaves. So, the potential thermal injury should be considered because very hot beverages are identified by the International Agency for Research on Cancer (IARC) monograph as Category 2A carcinogens for ESCC . Given tPublished studies related to hot tea consumption and EC risk were searched in the databases of Embase, PubMed, and Web of Science up to October 13, 2021. The search terms used were as follows: (a) “hot” or “high temperature,” (b) “esophageal cancer” or “esophageal carcinoma” or “esophageal neoplasm,” (c) “tea” or “beverage” or “drinking;” these searched keywords were combined with “and/or” without restrictions. In addition, we also retrieved papers in the reference lists and recent reviews.The studies included in the meta-analysis should satisfy all of the following criteria: (a) It was a case-control or cohort study; (b) The ORs or adjusted OR values and relevant corresponding 95% CIs were reported, or provided sufficient information to calculate it; (c) The study was about tea individually or tea was definitely included in the study; (d) The study tested the relationship between EC risk and tea temperature; (e) The diagnoses were confirmed as EC; and (f) It was published in English.The quality of included studies was evaluated by two reviewers independently with the Newcastle-Ottawa Scale (NOS) recommended by the Cochrane Non-Randomized Studies Methods Working Group . This waTwo researchers completed the selection process independently to review the eligibility of all the studies and extract the required information, including the first author's family name, year of publication, country, study design, type of cancer, gender of subjects, number of cases, and controls, temperatures of tea, ORs and 95% CIs, and adjusted confounders. Reduplicative references were deleted using the Endnote software, and then, we read the titles and abstracts to check whether the articles met the inclusion criteria. If it is difficult to determine the eligibility of one paper according to the title and abstract solely, the full text is downloaded and checked for the final decision. Any disagreements about research choices among researchers were resolved through discussion.Adjusted ORs were extracted preferentially to non-adjusted ones; however, unadjusted ORs and 95% CIs were accepted when adjusted ORs were not provided. The one with the most adjusted variables was selected when more than one adjusted OR was reported. Where multiple risk estimates were available in the same study, for example, studies providing ORs for both ESCC and EAC, they were included as two separate studies. Where different temperatures were reported, the highest qualitatively described temperature was chosen, such as “hot,” “scalding,” and “very hot”.p < 0.05. The heterogeneity was assessed with chi-square based on Cochran's Q statistic  version 2.0, like combing the effect size of ORs and 95% CIs, generating forest plots and funnel plots, and determining whether there was a statistical association. The statistical significance threshold was set at tatistic , and theogeneity . Generalion bias , 28.We retrieved 210 unique articles from the databases of PubMed, Embase, and Web of Science, 61 of which were identified as potentially relevant. After reviewing the full text, we determined that 23 reports met our inclusion criteria , 29–48. Finally, 23 articles  were identified in this meta-analysis including 5,050 cases and 10,609 controls. The detailed characteristics of the included studies are shown in The qualities of the studies included were all evaluated by the NOS method, and all eligible studies scored highly , so the random-effects model was selected. The overall results of this meta-analysis showed that people who drink hot tea have a higher risk of EC than those who do not drink hot tea, with a combined OR value of 1.77 , indicating that hot tea drinking can significantly increase the risk of EC (Significant heterogeneity was observed among the included studies (Isk of EC .p < 0.001) for ESCC, was 1.93  for combined EC. Five studies were included in the meta-analysis for EAC, comprising 680 cases and 1,313 controls. There was a statistically non-significant decreased risk of EAC in patients who consumed hot tea, with a pooled OR of.76 (95% CI = 0.58–1.01). There was no statistically heterogeneity  for EAC.When we stratified the included studies according to the cancer types, the results of ESCC and EC were consistent within the overall articles. The pooled OR was 2.33 , HB-CC studies , and PS-CH studies  were all consistent within the overall conclusions , Africa , and the Americas  were consistent with the overall conclusion, while the results of those conducted in Europe  and Australia  were the opposite  to help detect influential studies. Interestingly, when we omitted one study in turn, the ORs varied from 1.70 to 1.84, and the p-value was always < 0.001, which indicated that the overall result was robust .p = 1) and Begg's rank correlation test .Based on the visualization of the funnel plot , it was This meta-analysis assessed the relationship between hot tea drinking and EC risk, according to 23 published case-control and cohort studies. The overall result indicated that drinking hot tea could significantly increase the risk of EC. A recent study also confirmed it was the temperature effect, but not polycyclic aromatic hydrocarbons (PAH) exposure that posed an EC risk . The incSignificant heterogeneity was observed in the overall meta-analysis for EC, and we failed to identify any individual study that was an important contributor to the heterogeneity by omitting one study in turn in the sensitivity analysis.The subgroup analyses generated by stratifying the studies according to study design or sex were both consistent with the overall result. Publication bias was not present in both the ESCC and EAC subgroup analyses. However, the studies performed in Europe and Australia indicated a risk reduction trend, while in Asia, Africa, and America showed a significant trend of increased risk. The reason may be that EAC represents the majority of EC subtypes in these countries and this subtype is not easily influenced by hot temperature . FurtherNotably, the association between hot tea drinking and EC risk may be affected by other confounding factors. For example, in Western populations, heavy tobacco smoking and alcohol consumption are the main risk factors for ESCC , 53. TheIn previous studies, the tea temperature was mainly estimated by self-reported perception, the results may vary across individuals and could not be objectively verified. Besides, the existing evidence in humans for the carcinogenicity of drinking hot beverages is limited. According to the IARC classification system of carcinogens, “drinking very hot beverages at above 65°C” has been classified as “probably carcinogenic” (Group 2A), rather than “carcinogenic” to humans (Group 1) . Indeed,There were several limitations in this meta-analysis. Firstly, the heterogeneity could be eliminated, neither in overall results nor in most the subgroup analyses. For example, some studies only provided the unadjusted ORs, while others reported the adjusted ones. Moreover, the adjusted confounders were not always the same in different studies. All of these factors could explain the heterogeneity in the meta-analysis to some extent. Secondly, sufficient data were not provided in some literature. Lastly, most of the questionnaires used in the included studies were qualitative regarding the temperature of tea consumption, which was relied on self-reporting by the participants. Thus, we could not extract the exact temperature of tea drinking, and a more precise analysis could not be performed.This meta-analysis indicates that hot tea consumption is associated with a significantly increased risk in EC, particularly in ESCC. Given that hot tea consumption is prevalent in modern society, the results of our meta-analysis have important implications for EC etiology research as well as EC prevention.The original contributions presented in the study are included in the article/HL designed the research, performed the literature search and statistical analysis, interpreted the data, and drafted the manuscript. HG designed the research, performed the literature search, handled funding, and made critical revisions of the manuscript. All authors read and approved the final manuscript.This work was supported by the National Natural Science Foundation of China (no. 81773230) and the Science and Technology Research Plan of Henan (no. 212102310619).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.


However, this effect might be biased due to users’ definitions, exposure periods, and other confounding factors. Large epidemiological studies with controlled bias are therefore warranted to confirm or refute the association found in this study. Considering the possible carcinogenic effect of PPI, physicians should be vigilant when prescribing high-dose or long-term PPI.Proton pump inhibitors are commonly prescribed medications for gastrointestinal disorders, which bring gastric acid down to normal levels. However, the effects of PPI on pancreatic risk remain unclear. Therefore, we conducted a systematic review and meta-analysis to investigate the association between PPI use and pancreatic cancer. The overall combined estimate suggested that PPI therapy was significantly associated with an increased risk of pancreatic cancer . Subgroup analysis showed that the pooled RR for rabeprazole and lansoprazole was 4.08 (95%CI: 0.61–26.92) and 2.25 (95%CI: 0.83–6.07), respectively. Moreover, the risk of pancreatic cancer was established for both the Asian  and Western populations . The findings of this updated meta-analysis demonstrate that the use of PPI was associated with an increased risk of pancreatic cancer. Future studies are needed to improve the quality of evidence through better verification of PPI status , adjusting for possible confounders, and ensuring long-term follow-up.Previous epidemiological studies have shown that proton pump inhibitor (PPI) may modify the risk of pancreatic cancer. We conducted an updated systematic review and meta-analysis of observational studies assessing the effect of PPI on pancreatic cancer. PubMed, Embase, Scopus, and Web of Science were searched for studies published between 1 January 2000, and 1 May 2022. We only included studies that assessed exposure to PPI, reported pancreatic cancer outcomes, and provided effect sizes (hazard ratio or odds ratio) with 95% confidence intervals (CIs). We calculated an adjusted pooled risk ratio (RR) with 95%CIs using the random-effects model. Eleven studies (eight case–control and three cohorts) that reported 51,629 cases of pancreatic cancer were included. PPI was significantly associated with a 63% increased risk of pancreatic cancer (RR With an annual incidence of approximately 0.5 million, pancreatic cancer is the 12th most common cancer and the seventh most frequent cause of cancer death with >0.43 million deaths annually ,7,8. TheProton-pump inhibitors (PPIs) are first-line medications in the clinical practice for the management of acid-related disorders. Since PPIs are often considered safe over-the-counter medications; therefore, less attention is paid by healthcare providers. Nowadays, adverse effects of long-term use of PPIs are gaining increasing attention, especially in the risk of gastric , colorecTherefore, we conducted an updated systematic review and meta-analysis of existing observational studies that evaluated the association between PPI and the risk of developing pancreatic cancer.We followed the guidance provided by the Cochrane Handbook ; thus, oData Sources and Search Strategy: PubMed, Embase, Scopus, and Web of Science were searched for published studies related to PPIs and pancreatic cancer between 1 January 2000, and 1 May 2022. Search terms used included “proton-pump inhibitor(s)” OR “omeprazole” OR “pantoprazole” OR “lansoprazole” OR “esomeprazole” OR “rabeprazole” AND “pancreatic cancer” OR “neoplasm(s)” OR “pancreatic malignancy(ies)”. The titles and abstracts of retrieved studies were screened to exclude irrelevant studies. The full texts of the remaining studies were examined to extract information that evaluated the effects of PPI on pancreatic cancer risk. The reference lists of the retrieved studies were also examined to obtain additional studies.Inclusion and Exclusion Criteria: We included all observational (cohort or case–control) studies that assessed exposure to PPI and risk of pancreatic cancer. All the observational studies needed to be published in English and provide effect sizes (OR/HR) with 95%CIs. Included studies were also required to provide clear information regarding the patients’ characteristics, inclusion, and exclusion criteria. Studies were excluded if they were reviews, editorials, case-reports, or letters to editors without data description.Data Extraction: Two authors (TNP and MMI) developed screening guidelines and examined the appropriateness of all included studies for inclusion. The two authors extracted the following information from each study: (i) author’s first and last name, publication year, country of the participants; (ii) study design; (iii) number of participants, age, gender; (iv) inclusion and exclusion criteria, adjusted confounding factors; (v) definition of PPI exposure, long-term PPI use, individual PPI exposure; (vi) effect sizes (OR/HR), and 95%CIs.Quality Assessment: The same two authors independently examined the quality of all included studies using the Newcastle-Ottawa Scale (NOS) recommended by the Cochrane library [ library ,17. The Data Analysis: The random-effects model (DerSimonian and Laird) was used to calculate the pooled RR and 95% CI. The heterogeneity between studies was estimated using two different methods. First, we calculated Cochran’s Q statistic for assessing heterogeneity. It tests the null hypothesis that all included studies obtain the same underlying magnitude of effect. The Q statistical test usually has insufficient power to distinguish a moderate degree of heterogeneity [p-value of <0.05 was considered to indicate significant heterogeneity. Second, we also calculated the I2 statistic to determine the proportion of total variation between studies due to heterogeneity rather than chance. In this case, values of I2 of <25%, 25 ~ <50%, 50 ~ <75%, and >75% were categorized as null, low, moderate, and high heterogeneity, respectively [p ≤ 0.05). We also used the funnel plot of the logarithm of RRs versus their standard errors. All analyses were performed using the Comprehensive Meta-analysis Software (CMA) version V3 . All statistical tests were two-sided, and a p-value < 0.05 was considered to be statistically significant.ogeneity . A p-valectively ,20,21. WSearch Results: Our primary search selected a total of 482 studies. After reviewing the titles and abstracts of these 482 studies, 464 were excluded as ineligible as they were duplicates, reviews, case-reports, letters, and others which did not meet the prespecified inclusion criteria. The remaining 18 studies went through full-text evaluation. Of these, a further seven studies were excluded for reasons presented in Study Characteristics: The characteristics of these 11 studies are presented in Quality of Included Studies: Two cohort studies had a NOS score of 9, and the remaining cohort study had a score of 8. For the case–control studies, four out of eight studies (50%) were of low quality (NOS score < 7), with an average NOS score of 7.27.PPI Use and the Risk of Pancreatic Cancer: Among the included 11 studies, the use of PPI was associated with a statistically significant 63% increase in pancreatic cancer . There was, however, considerable heterogeneity across the studies . Subgroup Analysis: We conducted subgroup analyses of the included 11 studies based on study design, location, adjusted factors, number of participants, study quality, and individual PPIs use . The overall pooled analysis of the three cohort studies demonstrated a significant positive association with pancreatic cancer . However, the overall pooled analysis of individual PPI use showed a non-significant association with pancreatic cancer.PPIs use . The adjSensitivity Analysis: Since the overall findings had high heterogeneity , we performed a sensitivity analysis. In order to assess the overall impact of a single study on pancreatic cancer risk, a sensitivity analysis was conducted by excluding studies one by one. First, we excluded Bosetti et al.’s [p = 0.002, I2 = 99.23%). Second, Boursi et al.’s [p = 0.002, I2 = 99.31%). However, the sensitivity analysis did not substantially change the pooled effect and the level of heterogeneity.et al.’s  study fret al.’s  study waPublication Bias: We used Egger’s regression to detect overall publication bias and also generated Begg’s funnel plots (p = 0.55). We utilized Duval and Tweedie’s trim-and-fill methods, and the adjusted RR was 1.43 (95% CI: 1.40–1.47); hence, the impact of this bias was probably close to null. el plots . HoweverThis is an updated meta-analysis of eleven observational studies involving more than 1.5 million individuals, which evaluated the effect of PPI use on the risk of pancreatic cancer. Our meta-analysis found a moderately increased risk of pancreatic cancer in people using PPI. However, this association was not statistically significant when stratified by region.p < 0.001) but also high heterogeneity (I2 = 99%). This study also lacked subgroup analyses. Furthermore, Hong et al. [2 = 98.75%). Their study also did not provide subgroup analyses or any information regarding doses and duration. In contrast, our updated meta-analysis used a higher number of studies and conducted comprehensive subgroup analyses to assess the differential effects of PPI use on pancreatic cancer. Furthermore, our study showed the effect sizes with several confounding factors that were not addressed by previous studies.Previous meta-analyses of observational studies on PPI uses and the risk of pancreatic cancer reported that there was a positive link between them ,32,33, w12 absorption [12 plays a key role in pancreatic cancer as reported in previous studies [The etiology of pancreatic cancer is multifactorial; age, sex, geographical location, genetic and behavioral factors are key contributors ,35. Althsorption  because  studies ,43.In the subgroup analysis discerning between study populations, the risk of pancreatic cancer was higher in the Asian than in the Western population. Regional differences are always complicated. Previous evidence demonstrated that many physiological determinants, such as genetic factors ,45, lifeSubgroup analyses showed an insignificant association among the studies adjusted with smoking status. Previous evidence highlighted smoking as a recognized risk factor for pancreatic cancer ,51. An iDue to limited evidence and lack of high-quality studies on the effect of PPI use on pancreatic cancer, we believe that the following aspects are needed to be investigated in future studies and in the real-world clinical practice.(i) Study design: All the studies can use the Strengthening the Reporting of Observational Studies in Epidemiology (STROBE) guidelines  to follo(ii) PPI dose and duration: All studies may provide information regarding the short and long-term users and the risk of pancreatic cancer. For example, physicians often prescribe PPIs for 15 days to treat common symptoms of GERD. However, physicians also consider the long-term use of PPIs for patients with severe erosive esophagitis or Barrett’s esophagus. However, there is no appropriate formula on how to define long-term PPI use , and the(iii) Confounding bias: In the real-world clinical setting, randomized control trials are often considered a “gold standard” method because they control for potential risk bias. However, confounding factors/effects occur in observational studies and change the outcome of interests, either directly or indirectly. These biases can weaken or strengthen or alter the actual association. All included studies adjusted confounding factors except for two studies ,29. HoweLimitations: Our meta-analysis has several limitations. First, all included studies in this meta-analysis were observational studies. Observational studies are often susceptible to uncontrolled biases, even if they are well designed, which may weaken the actual quality of the analysis. Second, two-thirds of the included studies in our meta-analysis came from Western countries, and only three studies were from Asia, even though the prevalence and mortality of pancreatic cancer have been rapidly increasing in Asia [ in Asia ,49. Thir in Asia ,23 and d in Asia ,13,23,24We conducted an updated meta-analysis of the association between PPI use and pancreatic cancer risk in more than 1.5 million participants. The findings of this study show that PPI use was significantly associated with an increased risk of pancreatic cancer. Our robust analyses contribute to a better understanding of PPI use and pancreatic cancer risk. Since PPI is associated with an increased risk of pancreatic cancer, physicians should be cautious when prescribing PPI for GERD patients. More epidemiological and mechanistic studies are warranted to further examine the relationship between PPIs and the risk of pancreatic cancer. Future studies should also focus on adjusting possible confounders and identifying the probable mechanisms underlying this association.


Diphenhydramine is an H1-antihistamine that counteracts the toxic effects of organophosphate and carbamate insecticides that inhibit cholinesterase (ChE) activity. This meta-analysis aimed to investigate the effects of diphenhydramine on ChE inhibition induced by these insecticides in the plasma, erythrocytes, or whole brain of experimental animals.A data search was performed on erythrocyte, plasma, and brain ChE inhibition caused by organophosphate and carbamate insecticides in experimental animals  treated with the antihistamine diphenhydramine in accordance with preferred reporting items for systematic reviews and meta-analysis, which was done by the two-group random-effects model meta-analysis. The meta-analysis included 18 records extracted from six studies that, appeared from 1996 to 2022.2 of the combined effect size was 74%, with a significant Cochrane Q-test result . Subgroup analysis indicated that, with brain ChE inhibition, the heterogeneity (I2) became 5%, which was lower than ChE inhibition in plasma (84%) and erythrocytes (78%). No publication bias was identified using the funnel plot and Egger’s test.Using the random-effects model, a two-group meta-analysis revealed that the combined effect size (ChE inhibition) was significantly more favorable in the control group than in the diphenhydramine intervention, as shown by a forest plot. The combined effect size (standardized mean difference) was 0.67, with a standard error of 0.3, a lower limit of 0.04, and an upper limit of 1.29 (p = 0.025). The heterogeneity was moderate, as IThis meta-analysis suggests that, in addition to its documented antidotal action against ChE-inhibiting insecticides, diphenhydramine can also reduce the extent of ChE inhibition, especially in the brain, which is the main site of toxicity of these insecticides. There is a need for additional studies to assess such enzyme inhibition in different parts of the brain. In veterinary clinical practice, public health, and agriculture, there is a widespread use of formulations of organophosphate (OP) or carbamate (CA) insecticides to control pests –3. Such Diphenhydramine is an H1-antihistamine that counteracts the toxic effects of OP and CA insecticides , 10. Sevin vivo animal experiments, it was reported that diphenhydramine did not affect the extent of ChE inhibition in rats [in vitro, diphenhydramine partially reduced the inhibitory actions of the CA insecticide carbaryl on PChE and EChE in humans [in vivo in chicks [Regarding the mechanism behind the antidotal action of diphenhydramine against OP or CA insecticide, its antimuscarinic and possibly antinicotinic properties have been highlighted , 22, 23. in rats , 15 and  in rats , whereas in rats –22. It wn humans  and on Pn humans . Meanwhin chicks . These fn chicks , 22, 30,n chicks , 34.st time the effects of antihistamine on ChE inhibition induced by OP (dichlorvos) or CA  insecticides, as found in the literature.Against this background, and because of the various, although conflicting, studies on ChE activities under the influence of diphenhydramine with OP or CA insecticides, this meta-analysis was established to investigate for the 1This is a meta-analysis, which was approved by the Reviewing Board, College of Pharmacy, University of Duhok, Iraq, and no live animals were used.The study was conducted from January 2, 2022, to July 31, 2022, at the College of Pharmacy, University of Duhok, Iraq.The data search was based on research articles and academic theses published from 1980 to July 31, 2022, on the combined effects of diphenhydramine with OP or CA insecticides in experimental animals , using PubMed, Google Scholars, Scopus (Science Direct), Directory of Open Access Journals, Semantic Scholar, Web of Science, My Research Assistant, and Iraqi Academic Scientific Journals databases. A manual search was performed on articles and academic theses from Iraqi universities because most of these resources do not appear in the above-mentioned databases. No language restriction was applied to the search strategy.in vivo influence of OP or CA insecticides with concurrent diphenhydramine  or saline (control) treatment. The diphenhydramine treatments were administered either before or after the OP or CA insecticide dosing.As shown in In vitro studies were also excluded from the study.Studies included in the meta-analysis were those reporting effects of diphenhydramine on ChE activities in blood or tissue in experimental animals treated with OP or CA insecticide. Records were excluded if they were from studies using antihistamines other than diphenhydramine or did not report ChE activity in experimental animals. Data on ChE activities, as described in the selection criteria, were extracted from the texts, tables, or figures of the identified manuscripts. They reported ChE inhibition in experimental animals subjected to OP or CA insecticide treatments with or without diphenhydramine therapy. Some papers reported more than one subset of ChE inhibition data; such data were also included in the meta-analysis. We calculated the standard deviations from the corresponding standard errors depending on the data presented in the studies. When the numbers of experimental animals were presented as a range of minimum-maximum per group, the minimum number was chosen, to avoid increasing the favorable outcome toward the treatment effect. Any disagreement about data extraction was resolved by the authors of this paper through discussion.https://www.erim.eur.nl/research-support/meta-essentials/download/), which included a forest plot to calculate the effect size, weighted means with their 95% confidence intervals, a funnel plot, and subgroup analysis [https://www.meta-mar.com/) as further supporting evidence, with proposed confidence intervals at p < 0.05 [A two-group meta-analysis was conducted in accordance with PRISMA guidelines using the random-effects model because of the expected sources of heterogeneity among different studies  14, 15,–22, 31. analysis . Along wp < 0.05 .2 was calculated using the above-mentioned software , with values ranging from 0% (no heterogeneity) to 100% (high heterogeneity) [The Cochrane Q-test was used to statistically examine any heterogeneity –40 at a geneity) –40.A funnel plot (size effect against standard error) was used to visually assess any publication bias, while Egger’s statistical test was applied to objectively confirm the existence of such bias , 41.Using the random-effects model, a two-group meta-analysis of these records revealed that the combined effect size (ChE inhibition) was significantly more favorable in the control group, as shown by the forest plot, indicating the value of diphenhydramine for treating OP or CA poisoning . This va2 index to examine the heterogeneity among the studies of this random-effects model of meta-analysis . The drapery plot  provided2) became 5% , compared with equivalent PChE and EChE values of 84%  and 78% , respectively. Correspondingly, the weight of BChE in the subgroup analysis was 81%, compared with 12% for PChE and 7% for EChE.To identify the cause of heterogeneity, subgroup analysis was conducted. The related forest plot with the associated statistics is depicted in To examine whether there was publication bias, the funnel plot was visually examined. This indicated no publication bias, given the symmetrical distribution of the effect size points in a limited area of the plot . This coStudies in humans , 24 and This meta-analysis was conducted to resolve the potential beneficial effects of diphenhydramine regarding ChE inhibition. It demonstrated the beneficial effects of diphenhydramine in cases of poisoning by OP or CA insecticides in experimental animals at the level of ChE. This was shown by the combined effect size (high ChE inhibition) being significantly more favorable in the control-untreated (non-diphenhydramine) group; that is, lower ChE inhibition in the diphenhydramine intervention group . However2 index of 74%) in the results. Interestingly, subgroup analysis demonstrated no heterogeneity with BChE (I2 = 5%), while PChE and EChE had high heterogeneity . It is plausible that the heterogeneity was attributable to the inherent variations in ChE activities found in the plasma, erythrocytes, and brain [in vivo , in vitro studies on human PChE and EChE [This meta-analysis suggested significant heterogeneity (albeit with a moderate Ind brain –44. BChEnd brain , 42. Furnd brain . Furthernd brain . Howevernd brain , 44. In and EChE  and chicand EChE , 22 suggVisual examination of the funnel plot together with results of the objective Egger’s test indicated an absence of publication bias in the outcome covering the studies under this meta-analysis. In the contour-enhanced funnel plot, there was a symmetrical distribution of the effect size points . This obThis meta-analysis suggests that, in addition to its documented antidotal properties against poisoning with OP or CA insecticides, diphenhydramine could also reduce the extent of their ChE inhibition, especially in the brain, which is the main target of toxicity upon intoxication with these insecticides. There is a need for additional studies about the interaction of diphenhydramine with ChE-inhibiting insecticides in different parts of the brain, which are crucial in cases of poisoning with these insecticides.FKM: Conceptualized the study and drafted the manuscript. FKM, HMSG, AAM and HMR shared equally in the review and meta-analysis, provided comments, participated in writing the manuscript, and approved the final version of the manuscript.


N2 = 0.735). In addition, higher NFOGQ scores were associated with a more significant anteroposterior root mean square distance of the center of mass in the IMA direction and a greater inner stance phase in the outer step of the more affected limb (OMA) direction; explanatory power was 50.1%. Assessing the maximum speed and turning direction is useful for evaluating the differences in turning characteristics between freezers and non-freezers, which can help define freezers more accurately.For people with Parkinson’s disease (PD) with freezing of gait (FOG) (freezers), symptoms mainly exhibit as unilateral motor impairments that may cause difficulty during postural transitions such as turning during daily activities. We investigated the turning characteristics that distinguished freezers among people with PD and analyzed the association between the New Freezing of Gait Questionnaire (NFOGQ) scores and the gait characteristics according to the turning direction for the affected limbs of freezers. The study recruited 57 people with PD . All experiments measured the maximum 180° turning task with the “Off” medication state. Results revealed that the outer ankle range of motion in the direction of the inner step of the more affected limb (IMA) was identified to distinguish freezers and non-freezers (R Freezing of gait (FOG) is defined as a symptomatic episodic absence of forward progression or marked reduction in step length, despite the intention to walk , and appThe New Freezing of Gait Questionnaire (NFOGQ) checks for the presence or absence of FOG and objectively evaluates the severity . In addiFreezers have higher cadence, increased step time variability, and disordered bilateral coordination during turning, deficits that correlate with a greater number of FOG episodes ,14. FurtThis study aimed to distinguish between freezers and non-freezers based on 180° turning characteristics and to investigate the associations between NFOGQ scores, which determine the presence and severity of the FOG and 180° turning characteristics at the maximum speed. According to our hypotheses, the 180° turning characteristics based on a full-body kinematic analysis may distinguish freezers from non-freezers and have a strong association with the NFOGQ score.A total of 125 people with PD were recruited at the Department of Neurology of a local medical center. People who did not meet the inclusion criteria and refused to participate were excluded, and 73 people with PD were allocated to the groups (freezers and non-freezers). The drop-out criteria were as follows: (1) those who withdrew their informed consent; (2) those who could not attend the entire experiment due to severe symptoms; (3) and those who showed FOG symptoms in the turning phase. Therefore, this study conducted the final analysis with 57 people with PD . Their condition was diagnosed by a neurologist according to the UK Parkinson’s Disease Society Brain Bank criteria . InclusiOne participant had a turning task-induced FOG episode with severe symptoms difficult to analyze. Therefore, data for this participant was excluded.The study flow chart is shown in Participants performed two sessions. In the first session, the participants completed the informed consent form and were assessed using the UPDRS, modified Hoehn and Yahr scale, NFOGQ, and MMSE . In the All participants wore fitted T-shirts and shorts for convenience, and the turning task was performed barefoot. First, body height, body weight, shoulder offset, elbow width, wrist width, hand thickness, leg length, knee width, and ankle width measurements were obtained to estimate the joint kinematics data. Then, 39 reflective markers in the shape of spheres (14 mm each) were placed according to the Plug-in Gait full-body model  .The participants warmed up by walking and stretching for approximately 5 min before the experiment started. Then, the participants were instructed to practice the 180° turning task at the maximum speed with 3 to 5 trials, and the measurements were conducted after approximately 5 min of rest. After the initial rest period, the participants were instructed to sit on a chair. At the beginning signal using the bell, the participants were asked to rise from sitting on a standard chair, walk 3 m, turn around a cone, walk back to the chair, and sit down at maximum speed. The 180° turning task at the maximum speed used in this study was modified from the timed up and go test a 11]..11].The 180° turning task at maximum speed was performed in the IMA and OMA directions, three times each in random order. The direction of the more affected limb was defined as the asymmetry level in the motor symptoms of the participants using the items that individually evaluated the physical function of the UPDRS, and the pathological symptoms were more dominantly observed . A neuroX-axis, Y-axis, and Z-axis faced the right, anteriorly, and superiorly, respectively. In addition, the Z-axis was defined as the cross-product between the X-axis and Y-axis. The collected data were sampled at 100 Hz and filtered using a fourth-order Butterworth low-pass filter with a 10 Hz cutoff frequency ..9].The analysis variables, spatiotemporal, kinematic, and center of mass (COM) parameters during the 180° turning tasks at maximum speed were calculated using MATLAB R2020b . An average of three trials was used in the analysis. The definitions of the analysis variables are presented in At study commencement, we performed a priori power analysis. This analysis showed that a minimum sample size of 56 participants was required to achieve a statistical significance of 0.05 with a power of 80% at an overall effect size of 0.20 when set to 2 for the number of groups and 4 in the number of measurements. Furthermore, considering the rate of dropouts that may occur during the experiment, the number of participants was set as 33 freezers and 33 non-freezers ,24. The U test  or independent t-test  were conducted to assess the differences in physical and clinical characteristics between freezers and non-freezers. In addition, a two-way analysis of variance with repeated measures was performed to investigate the main effects and interactions of freezers and non-freezers and between the turning directions (IMA and OMA directions) of the 180° turning tasks at the maximum speed. Finally, the post hoc test was conducted using the analysis of covariance (ANCOVA) or the Mann-Whitney U test between freezers and non-freezers. The selected covariates were age, sex, height, and body mass index (BMI) in ANCOVA to minimize the influence of these factors affecting the participants’ demographic characteristics. Additionally, the effect size was calculated using Cohen’s d. This study aimed to identify variables that can distinguish groups, so we focused on the results comparing the groups.All statistical analyses were performed using SPSS 21.0 , and the statistical significance level was set at 0.05. The Shapiro–Wilk test was used to assess data normality. Fisher’s exact test and Mann-Whitney After Z standardization was performed with the turning characteristics that showed differences between groups, a stepwise binary logistic regression analysis was performed to identify classifier variables for distinguishing freezers and non-freezers. The covariates and turning characteristics set for the first and second blocks were age, sex, height, BMI, and differences between freezers and non-freezers . The classifier variable was expressed as an odds ratio with a 95% confidence interval (CI). In addition, the area under the curve (AUC) was calculated through the turning characteristic’s receiver operating characteristic (ROC) curve analysis of the turning characteristic to identify the classification accuracy of freezers and non-freezers. Finally, Youden’s index was used to analyze the optimal cutoff value of the variable that distinguished the freezers from non-freezers. AUC > 0.9 indicates high accuracy, whereas AUCs of 0.7–0.9 and 0.5–0.7 indicate moderate and low accuracies, respectively . A partip = 0.031), outer double support phase (p = 0.036), and lower outer ankle range of motion (ROM) (p = 0.048) when compared with non-freezers in the IMA direction. In addition, freezers demonstrated significantly lower inner hip ROM (p = 0.049) than non-freezers in the OMA direction. All characteristics analyzed during the 180° turning task are summarized in Post hoc analyses revealed an effect of groups; the differences in 180° turning characteristics freezers had significantly greater total steps  in the IMA direction, and the explanatory power was 73.5% (RN2 = 0.735).The results of the stepwise binary logistic regression and ROC curve analysis  showed tp = 0.024) and a significant positive correlation with anteroposterior (AP) root mean square (RMS) distance of the COM  in the IMA direction. In the OMA direction, the NFOGQ score had a significant positive correlation with inner stance phase . The variables in the results of the correlation analysis are presented in We first assessed the correlation between NFOGQ scores and turning characteristics in freezers. The NFOGQ score showed a significant negative correlation with inner toe clearance height  in the IMA direction and greater inner stance phase (p = 0.018) in the OMA direction (2 = 0.501). This is the linear regression equation:Then, we further analyzed the associations between NFOGQ scores and turning characteristics using a stepwise multivariable linear regression model after adjusting for age, sex, height, and BMI. The results for freezers showed that the higher NFOGQ score was associated with greater AP RMS distance of the COM  between the caudate and putamen nuclei—the areas responsible for motor performance control —not onlyThese asymmetric characteristics of freezers affect inter-limb coordination and decrease the ability to posture control during movement due to posture instability in daily life . ConsideWe showed that several 180° turning characteristics were related to the NFOGQ scores in freezers, and the results of the linear regression analysis model were able to further establish the association through a robust explanatory power of 50.1%. In previous studies, as NFOGQ scores in freezers increased, variables with a relationship between gait characteristics were reported, such as reduced step length, decreased mediolateral (ML) center of pressure, and increased gait asymmetry ,18. HoweFurthermore, freezers require a more dynamic balance than straight gait due to increased activities of the frontal cortex and basal ganglia circuits during turning, and FOG symptoms may occur due to this increased activity . Our resMancini et al.  reportedOur study has several important clinical implications. The main results of this study not only distinguished the turning characteristics of freezers among people with PD but also revealed the association between the NFOGQ scores and the gait characteristics based on the turning direction of the affected limbs of freezers. In addition, we found meaningful variables by considering clinical characteristics such as disease duration, levodopa equivalent dose, Hoehn and Yahr scale, UPDRS total scores, and UPDRS Part II scores in this analysis process for distinguishing freezers and non-freezers. We suggest that these factors need to be considered in future research designs. These approaches have an expected effect on improving the diagnosis of freezers, a severe stage affecting people with PD, and could be used as basic data for intervention, injury prevention, and rehabilitation education. We recommend that all stakeholders, such as patients, caregivers, healthcare providers, and clinicians, pay attention to turn movements of daily life for rehabilitation using turning tasks of freezers.p = 0.655) between freezers and non-freezers despite the significant difference in the UPDRS total scores. Although the UPDRS Part Ⅲ score may contribute to evaluating the functional effect of FOG, it does not reflect the overall severity of FOG [There are several limitations to this study. First, we did not compare the difference between the “On” and “Off” medication states while evaluating the 180° turning task. Dopaminergic medications affect gait characteristics in people with PD. Second, our results showed no significant difference in UPDRS Part Ⅲ scores (y of FOG ,31. Lasty of FOG . Since ty of FOG . In addiy of FOG . Understy of FOG ,34. TherThis study showed a significant difference between freezers and non-freezers in a kinematic variable during the 180° turning task at maximum speed in the IMA direction. In addition, lower turning performance might indicate increased FOG severity, which had a high explanatory power of 50.1%, such as the associations between NFOGQ scores and turning characteristics. Understanding the turning characteristics regarding maximum speed and turning direction in freezers and non-freezers will be useful for evaluating differences in turning characteristics, improving disease monitoring, and future clinical evaluations. Furthermore, our study results will help to define freezers more accurately, recognize the relationship between disease progression and impaired turning patterns, and improve the accuracy of the FOG prediction model.


Type 2 diabetes (T2D) associated health disparities among different ethnicities have long been known. Ethnic variations also exist in T2D related comorbidities including insulin resistance, vascular complications and drug response. Genetic heterogeneity, dietary patterns, nutrient metabolism and gut microbiome composition attribute to ethnic disparities in both manifestation and progression of T2D. These factors differentially regulate the rate of metabolism and metabolic health. Metabolomics studies have indicated significant differences in carbohydrate, lipid and amino acid metabolism among ethnicities. Interestingly, genetic variations regulating lipid and amino acid metabolism might also contribute to inter-ethnic differences in T2D. Comprehensive and comparative metabolomics analysis between ethnicities might help to design personalized dietary regimen and newer therapeutic strategies. In the present review, we explore population based metabolomics data to identify inter-ethnic differences in metabolites and discuss how (a) genetic variations, (b) dietary patterns and (c) microbiome composition may attribute for such differences in T2D. The convergence of biological/genetic factors associated with geographical origins, culture, economic, political/legal factors and race contributes to the multidimensional nature of ethnic diversity  cut-offs in different ethnic groups enabling better clinical management and further reducing the prevalence of T2D  comparing metabolic changes among ethnicities; (b) examining how these tissue-specific metabolites/metabolic pathways might contribute to vascular complications, differ between ethnicities and (c) investigating the influence of genetic constitution and dietary patterns on these consequent metabolic changes. Various factors contributing to pathological differences of T2D among ethnicities are summarized in Fig. Metabolomics involves comprehensive analysis of small molecules (< 1.5 kDa) providing both qualitative and quantitative measurements in biological systems  progressed to T2D over 12 years. Further, targeted MS analysis involving participants from a nested case-control study carried out in the identical cohort who progressed towards developing T2D pointed out a significant association between 2-aminoadipic acid and risk for T2D , alanine (1.63-fold) and oleic acid (1.87-fold) were positively correlated to T2D. On the other hand, lysophosphatidylcholines (LPC) 18:0 (20%) and creatinine (37%) showed a negative association with T2D  and the Shanghai Men’s Health Study (SMHS) revealed significant modulation of metabolites associated with pathways regulating glucose homeostasis (glycolysis/gluconeogenesis), BCAA, fatty acids, glycerophospholipids, androgen and bradykinin  aimed to understand metabolic status in T2D among various ethnic groups. The participants were Europeans  and South Asians  . The study demonstrated the relationship between circulating amino acids and the onset of T2D. Predisposing factors of T2D such as alcohol consumption and smoking were less in the South Asian ethnic groups. Isoleucine, tyrosine, alanine and phenylalanine levels were significantly elevated in the South Asian cohort. However, these amino acids showed cross-sectional correlations with glycemic index and insulin resistance in both ethnic groups. After a 19 year follow-up, the metabolite profile showed a strong association of tyrosine with the incidence of T2D in South Asians  associated with T2D. These loci are associated with gastric inhibitory peptide (GIP), major histocompatibility complex, class II, DQ alpha 1 (HLA-DQA1), neurexin 3 (NRXN3), C-Maf Inducing Protein (CMIP) and glucagon Like Peptide 2 Receptor (GLP2R) genes  gene which plays a role in pancreatic hypoplasia and neonatal diabetes  (rs10811661), CDK5 regulatory associated protein 1-like 1 (CDKAL1) (rs7756992), solute carrier family 30 member 8 (SLC30A8) (rs1326663), insulin like growth factor 2 mRNA binding protein 2 (IGF2BP2) (rs4402960), C2 calcium dependent domain containing 4 A (C2CD4A/B) (rs7172432), potassium voltage-gated channel subfamily Q member 1 (KCNQ1) (rs2237892), insulin receptor substrate 1 (IRS-1) (rs2943641), glucokinase regulator (GCKR) (rs780094) and potassium inwardly rectifying channel subfamily J member 11 (KCNJ11) (rs5219) were associated with predisposition to T2D. Further, Iwata et al., demonstrated a significant association between genetic risk score (P = 5.9 × 10− 21) and early onset of T2D  at 2q21 locus which is close to transmembrane protein 163 (TMEM163) is associated with reduced levels of fasting plasma insulin  belonging to the zinc transporter family is involved in insulin signalling. A meta-analysis was conducted to find the association between SLC30A8 polymorphism (rs13266634) and the risk of T2D. The study demonstrated that Europeans  1.11–1.18, p < 0.001) and Asians  harbouring this single nucleotide polymorphism (SNPs) are at increased risk of developing T2D  and Hispanics (21.9%) in contrast to the non-Hispanic White group. The study also stated that African-Americans (14.1%) and Hispanics (11.7%) showed a high prevalence of T2D when compared with non-Hispanic Whites  have indicated ethnicity based genetic markers associated with a variety of diseases including T2D Ali, . A meta-ce Guan, . A studyETFDH), monocarboxylate transporter 9 (SLC16A9), correspond to genes encoding solute carriers. ACADS, ACADM and ACADL genes are associated with beta (β)-oxidation of fatty acids. Fatty acid desaturase 1 (FADS1), and elongation of very-long-chain fatty acids-like 2 (ELOVL2) are linked with the biogenesis of polyunsaturated fatty acids and serine palmitoyl transferase long chain base subunit 3 (SPTLC3) is related to phospholipid biosynthesis  alters the rate-limiting step of fatty acid biosynthesis mediated by stearoyl-CoA desaturase , for heterozygous , under allelic are , for co-dominant , for recessive , for additive  and for dominant . East Asians had an co-dominant  and under allelic  variant   which falls in the category of calcium-dependent phospholipases (cPLA2s) and are involved in the cleaving of arachidonic acid from phospholipid membranes. Wolford et al., identified a C to G variant associated with PLA2G4A in the Pima Indians, which leads to the substitution of phenylalanine with leucine  (rs12231737) was associated with males . One of the significant missense variants associated with rs12231737 (Phet=2.6 × 10 − 19) is rs671  = 77.7%, 95% CI 1.14–1.20, Pmales=1.5 × 10 − 24). This variant is known to reduce ALDH2 activity  and cardiovascular risk in East Asians  gene. The study participants were a part of two different cohorts; the B99 cohort and Inuit Health in Transition (IHIT) located in Greenland. One of the variants was known to destroy a splice-acceptor site in exon 14 of ADCY3 encoding adenylate cyclase. This catalyses the formation of cyclic adenosine monophosphate (cAMP) from adenosine triphosphate (ATP). The study showed that loss-of-function variants in Greenlandic homozygous (ADCY3 c.2433-1G > A) resulting in a phenotype characterized by truncal adiposity and metabolic alterations associated with insulin resistance, dyslipidemia and T2D . Subjects with these variants showed higher levels of plasma HDL. Perilipin-1 in humans is vital for the formation of lipid droplets, triglyceride storage and metabolism of FFAs  downstream variants were linked with triglycerides and HDL levels and imply the association of this gene with T2D and metabolic syndrome  is correlated with serum levels of adiponectin in most of the studies  is linked with adiponectin levels   gene rs10889335, which is linked with phosphatidylinositol 1-stearoyl-2-arachidonoyl-GPI (18:0/20:3). DOCK7 is co-localized with low levels of angiopoietin-like protein 3 (ANGPTL3) in the liver. ANGPTL3 is associated with lipid metabolism and is known to cause type 2 hypobetalipoproteinaemia. Variants of TRIB1  are associated with LDL, CHD and HDL in Asian and European populations  which belongs to phosphatidylcholine species. These lipid molecules play a vital role in insulin resistance  are associated with reduced HbA1c levels. This variant has been significantly associated with the Malay population  (PPM1K) gene which activates Branched Chain Amino Acid ketoacid dehydrogenase (BCKD) is associated with T2D  Asian Genetic Epidemiology Network (AGEN) and (b) Diabetes Meta-analysis of Trans-ethnic Association Studies (DIAMANTE) with 77,418 T2D subjects and 356,122 controls revealed 301 distinct signals associated with 183 loci in T2D subjects. When gender based stratification was performed aldehyde dehydrogenase  trial have revealed that decreased caloric intake reduced the levels of HbA1c further minimized cardiovascular risk factors  is a gut metabolite produced as a result of intestinal microbial metabolism  such as propanoic acid, butyric acid and In this present review manuscript, we explored serum/plasma metabolomics data of T2D subjects in different populations reported in various studies and identified metabolites/metabolic pathways uniquely altered among ethnic groups. Further, we discussed that these metabolic differences may be due to genetic composition along with specific dietary intake reflecting in gut microbiome content among ethnicities. However, this is an observational study and no statistical and bioinformatics analysis was performed due to the unavailability of raw data for T2D metabolomics studies from different ethnic groups. Metabolomics data from diverse studies that we analysed were obtained from different platforms such as LC-MS, GC-MS and NMR and these techniques possess limitations on biochemical analysis of metabolites. Studies conducted to compare metabolic differences in different ethnicities had varied sample size and majority of studies were cross-sectional and case-control studies. The analysis did not consider extrinsic and intrinsic parameters such as BMI, gender, diet, lifestyle and ethnicity. Comparative metabolomics analysis among different populations of T2D, cited in this manuscript were performed in the non-natives from different geographical regions adapted to the newer environment which may mask effects of natural habitats/dietary patterns and lifestyle. Hence, an appropriate study may involve recruiting subjects from their natives and perform comparative metabolomics analysis in different ethnicities using one single platform as a consortium study.International Diabetes Federation. IDF Diabetes Atlas, 10th edn. Brussels, Belgium: International Diabetes Federation T2D is a complex metabolic disorder characterized by hyperglycaemia, insulin resistance or impaired insulin secretion or both. The 2021 reports of the International Diabetes Federation reveals 537 million adults (20–79 years) living with diabetes across the globe and these numbers may reach 783 million by 2045. The reports also state that three in four adults living with diabetes are from low-and middle-income countries. Epidemiological data indicates geographical differences in prevalence of T2D across the globe  identify metabolite correlates of extrinsic and intrinsic factors in large scale metabolomics analysis among various ethnicities and (b) prospective studies in individual patients over the time to get mechanistic insights to understand metabolic changes in T2D subjects during the progression of the disease to vascular complications. Moreover, it is known that the extent to which these metabolic processes may (dys)function differs between patients and hence, T2D must be treated as a self-management disease, focusing on prevention and cure instead of fighting symptoms. Identifying the