ADIPOQ rs2241766 gene SNP and breast cancer risk.We aimed to conduct a meta-analysis to accurately evaluate the potential association between A systematic literature search on Cochrane Library, PubMed, Embase, Web of Science and China National Knowledge Infrastructure (CNKI) identified 8 articles with 1692 cases and 1890 controls. Strength of association was evaluated by pooled odds ratio (OR), 95 % confidence interval (CI) and p value. Funnel plots and Begger’s regression test were applied for testing the publication bias. Statistical analysis of all data was performed by Stata 12.0.ADIPOQ rs2241766 gene polymorphism did not significantly associated with the risk of breast cancer for these genetic models .The meta-analysis results indicated that the ADIPOQ rs2241766 SNP and breast cancer. Further large-scale and well-designed studies will be indispensable to confirm our result.This study indicated that no significant relationship between the  Breast cancer was the most common malignant tumor in women, which was the second most common malignant tumor worldwide . AccordiADIPOQ) was inversely related to breast cancer and other tumors [ADIPOQ gene was found on the 3q27 chromosomal expressed by adipose tissue and had more than 620 variants [ADIPOQ gene polymorphism was closely related to cancer risk by influencing plasma level of ADIPOQ [ADIPOQ rs2241766 gene and breast cancer susceptibility were limited and rather contradictory [ADIPOQ rs2241766 gene polymorphism and breast cancer risk, which would have much greater possibility to reach reasonably reliable conclusions.Recent studies had shown that adiponectin (r tumors –6. ADIPOvariants . ADIPOQ f ADIPOQ . To dateadictory –17. For ADIPOQ” OR “adiponectin receptor” OR “rs2241766”) AND (“variant” OR “polymorphism” OR “mutation”) AND (“breast cancer” OR “breast tumor”). Two investigators manually checked the reference of retrieved articles and extracted the publications independently. In addition, only English and Chinese articles were included.We systematically searched on database of Cochrane Library, Pubmed, Embase, Web of Science and China National Knowledge Infrastructure, up to October 31, 2019, using the following terms:  ADIPOQ rs2241766 genotype was obtained. Main exclusion criteria as followed: (1) repeat of other articles; (2) not case–control studies; (3) unpublished studies, conference articles, meta-analysis and systematic evaluations; (4) pedigree studies. Consulting the Preferred Reporting Project (PRISMA) Guide for Systematic Evaluation and Meta-Analysis[All selected studies complied with the inclusion criteria: (1) full text can be found; (2) case-control studies focused on relationship between the -Analysis, by screADIPOQ rs2241766 genotype. Only the article with maximum sample size was selected while same data appearing in multiple publications. To insure the data accuracy, a third investigator reviewed the final results. Discussions were executed to solve disagreements.Two investigators extracted data independently according to the selection criteria. The following items were collected: first author, country, publication year, amount of cases and controls, Hardy-Weinberg equilibrium, control group source and the availability of Two researchers performed independent quality assessment for each eligible article according to 9-point Newcastle-Ottawa Scale (NOS) applied to quality evaluation of observational studies . DiffereADIPOQ rs2241766 gene SNP and the risk of breast cancer. The Chi-square based Q-test and I-squared test was used to analyze the heterogeneity (P<0.10 suggested Heterogeneity) [Odds ratio (OR) and 95 % confidence interval (CI) were calculated to estimate the relationship between geneity) , 21. Thegeneity) . In contgeneity) . Stata 1Flow diagram for the retrieve strategy was demonstrated in Fig. p<0.001, I2 = 84.2 %; T vs. G : p<0.001, I2= 84.3 %). However, the heterogeneity was not considered significant in these models (TT + TG vs. GG (dominant model): p=0.308, I2 = 16.0 %) Fig. ADIPOQ rs2241766 polymorphism and breast cancer risk under the model of TT vs. TG + GG , TT + TG vs. GG , C vs. T . We also performed a subgroup analysis according to ethnicity, source of control, case size and control size. We find the statistically significant association for the dominant model which observed in Asian population , control from population  and studies with less or more than 100 controls , and less or more than 100 cases. The pooled OR and 95 % CI of the relationship between ADIPOQ rs2241766 polymorphism and breast cancer were in Table p=0.458; TT + TG vs. GG: p=0.881; T vs. G: p=0.322) . Al Khaldi et al. [ADIPOQ rs2241766 gene SNP and increased risk of breast cancer [In recent years, several case–control studies have shown an association between the importance of  in 2008 , in this in 2008  found thi et al.  found tht cancer , 13–15.ADIPOQ rs2241766 polymorphism was not significantly associated with breast cancer risk. Considerable heterogeneity was discovered among these studies. We also conducted a subgroup analysis according to racial classification, case size, control size and source of control. In dominant genetic model, the significant correlation was not found in stratification analysis by racial classification, case size, control size and source of control. But in Asians, the statistically significant association for the dominant model was observed , control from population  and studies with less than 100 controls . Moreover, the significant associations were observed in allele genetic model in stratified analysis among Asians  and studies with less than 100 controls . Because of the limited samples and finite studies [ADIPOQ rs2241766 SNP has no relationship with breast cancer. New further discoveries will appear and the information will be constantly renovated. There should be plenty of research data to support more objective results.In the recessive model , dominant model  and allele model , the  studies , 12, it The detailed mechanism of the association between ADIPOQ rs2241766 and breast cancer is unclear. Previous epidemiological studies , 27 haveADIPOQ rs2241766 SNP. Finally, in all genetic models, the heterogeneity of the recessive model and allele model may have a significant impact on the result of meta-analysis.Several limitations in the study should be considered. First, only 8 studies were selected into our meta-analysis. The sample size and amount of studies were comparatively small, which affected the reliability of the results. Second, the case size and control size of some studies were relatively small to confirm the risk of ADIPOQ rs2241766 SNP could not be confirmed due to the relatively limited sample and small amount of study. Scholars had conducted extensive research on the susceptibility factors of breast cancer, which believed that the breast cancer occurrence was influenced by multiple factors [In conclusion, we find the significant associations in the recessive model and allele genetic model in stratification analysis. The risk of  factors –33, espe


Objectives: Invasive fungal infection (IFI) remains an important cause of mortality in liver transplantation (LT). The objective of this meta-analysis was to identify the risk factors for IFI after LT.Methods: We searched for relevant studies published up to June 2020 from PubMed, Web of Science, Embase, and the Cochrane Library. Odds ratios (ORs) and their corresponding 95% CIs were used to identify significant differences in the risk factors. Heterogeneity between studies was evaluated by the I2 test, and potential publication bias was assessed with Egger's test. The quality of included studies was evaluated with the Newcastle-Ottawa Scale (NOS).Results: A total of 14 studies enrolling 4,284 recipients were included in the meta-analysis. Reoperation , posttransplantation dialysis , bacterial infection , live donor , retransplantation , and fungal colonization  were associated with the risk factors of IFI after LT.Conclusions: Despite some risk factors that have been identified as significant factors for IFI post-LT, which may inform prevention recommendations, rigorous and well-designed studies with adequate sample sizes should be conducted to solve the limitations of this study. Since the first liver transplantation (LT) by Thomas Starzl in 1963, the prognosis was significantly improved. LT can be considered the most effective therapy for nearly all causes of end-stage liver diseases (ESLDs) . IFIs ocCandida species are the most common causes of IFI post-LT, followed by Aspergillus species  statement . The incThe following were the exclusion criteria: (1) duplicate articles; (2) reviews, meeting abstracts, letters, or case reports; (3) no diagnostic or no defined criteria for IFIs; (4) studies were related to the risk factors of IFIs after the organ transplantation but did not report the relevant data on the LT subgroup; (5) studies included data of risk factors for LT infection but did not show the information for IFI.Relevant information was extracted independently by the two reviewers (ML and LYS). A final check was confirmed by another researcher (ZJZ). The extracted data included the first author, publication year, country of origin, study time and design, number of patients and controls, and risk factors for IFIs with odds ratio (OR) and 95% CI from the multivariate analyses.The quality of each included study was independently evaluated by the three reviewers  based on the Newcastle–Ottawa Scale (NOS) . The NOSI2 > 50% or P < 0.1 (I2 < 50% or P > 0.05). Otherwise, the random-effects model was used. Sensitivity analysis was conducted by excluding one study at a time. Publication bias was assessed with Egger's test. P < 0.05 suggested that there was a publication bias in this study. The population attributable risk proportion (PARP) was calculated as follows: PARP = Pe (OR−1)/[Pe (OR−1) + 1]. Pe was defined as the pooled exposure rate of controls that represented the overall population exposure rate.Statistical analysis was performed using Review Manager 5.3 and Stata 12.0. Heterogeneity was considered significant with  P < 0.1 . The fixThe literature search yielded 2,405 results; most studies were excluded because they were duplicate studies or because they were not relevant to our meta-analysis. Then, 84 studies were excluded after the full-text articles were reviewed because they did not match the criteria. Finally, 14 studies involving 4,284 recipients  were included in this meta-analysis , 18–27. The specific characteristics of the studies included in the meta-analysis are presented in The risk factors for IFI post-LT are shown in Population attributable risk proportion was used to represent the proportion of cases in a population that was attributable to the exposed factor. The PARP of risk factors such as reoperation, posttransplant dialysis, bacterial infection, live donor, fungal colonization, and retransplantation is shown in Sensitivity analysis was used to evaluate the potential effect of heterogeneity conducted by eliminating one study in each turn. Sensitivity analyses manifested no significant changing of heterogeneity when one study was eliminated at a time.P < 0.05. Egger's test showed that most risk factors did not have a publication bias (P > 0.05).Egger's test was conducted for statistical investigation to evaluate potential publication bias . The pubInvasive fungal infection is associated with poor outcomes in recipients with posttransplant . TherefoCandida species were the most common causative organism of IFI among recipients of the LT, and Aspergillus species was the second most common.According to the inclusion and exclusion criteria, we identified 14 studies enrolling 4,284 patients. In our meta-analysis, risk factors for IFI after LT included reoperation, posttransplant dialysis, bacterial infection, retransplantation, live donor, and fungal colonization. As summarized in Reoperation and retransplantation, which indicated a more complicated intraoperative and postoperative procedure, were risk factors for IFI. Meta-analysis showed that the risk of IFI in recipients with reoperation was 2.18 times higher than the recipients without reoperation, which was consistent with the results of previous studies , 30. HowRecipients of the live donor were at a higher risk of IFI after LT, which was consistent with the previous studies , 31. In Our study also found that the risk of IFI increased with the posttransplant dialysis that is consistent with the studies of Ohkubo et al. , LavezzoWe used PARP to estimate the percentage of IFI in recipients of LT attributed to one kind of risk factor. We found that the PARP of the bacterial infection and reoperation was high. Thus, we infer that these risk factors were important for IFI post-LT.Fungal colonization was defined as the presence of fungus before LT without clinical symptoms or evidence of infection. We found that the patients with fungal colonization were at a higher risk of IFI. Several studies have shown that antifungal prophylaxis dramatically reduces fungal colonization, mortality caused by a fungal infection, and the overall incidence of fungal infection , 32. FurThere were several limitations to this meta-analysis. First, we only included English language literature from four databases, and there may have been incomplete retrieval. Second, there may have been mistakes in the data conversion because some study data required to be recalculated. Third, because of the limitations of the included date, we did not conduct subgroup analyses and funnel plots. At last, analyses were limited by the sample size included in this meta-analysis, so the combined effect size may have been exaggerated to draw the opposite conclusion.In conclusion, this meta-analysis identified some risk factors for IFI post-LT and might provide a basis for clinical prevention. However, a well-designed prospective cohort study should be conducted to validate our findings.The original contributions presented in the study are included in the article/supplementary material, further inquiries can be directed to the corresponding author/s.ML and LYS designed the study and wrote the first draft of the manuscript. ML and ZJZ verified data extraction, data analysis, and reviewed the manuscript. LYS and ZJZ supervised the data acquisition, data analysis, and interpretation. All the authors read and approved the final manuscript.This research was supported by the Beijing Municipal Science & Technology Commission (No. Z181100001718220).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.


Numerous dietary components have been linked to the development of islet autoimmunity (IA) and type 1 diabetes (T1D); however, no associations are firmly established. This systematic review and meta-analysis aimed to synthesize current knowledge on diet and incidence of IA and T1D.Literature search was performed in Medline, Embase, and Cochrane Library, from inception until October 2020. Eligible studies had IA or T1D as outcome; any dietary exposure; case-control, cohort, or randomized controlled trial design; and hazard, risk, or odds ratios as measures of association. Summary relative risks (RR) and 95% confidence intervals (CI) were estimated with random-effects models. Certainty of evidence was assessed with GRADE. PROSPERO registration number: CRD42020212505.2=43%) and exclusive  breastfeeding, later introduction to gluten , cow's milk , and fruit . Higher childhood intake of cow's milk was associated with increased risk of both IA  and T1D . For the remaining dietary factors investigated, there was no association, or the evidence was of low certainty.Among 5935 identified records, 96 were eligible, and pooled estimates could be produced for 26 dietary factors. Evidence with moderate/high certainty indicated lower risk of T1D in relation to longer , Novo Nordisk Foundation, and Swedish Diabetes Foundation. Certainty of evidence was evaluated for each association.Over the past decades, several dietary exposures at different developmental stages have been assessed in relation to type 1 diabetes (T1D), and many of those have been proposed as potential risk factors. Previous meta-analyses assess a limited number of dietary factors and ours is the first attempt to synthesize the totality of evidence regarding the influence of diet on T1D risk. We searched Medline (Ovid), Embase, and Cochrane Library (Wiley) for cohort and case-control studies, as well as randomized controlled trials (RCTs), published until October 15It was possible to synthesize data for 26 dietary exposures in utero, infancy, childhood, and/or adulthood in relation to IA and T1D from 96 observational studies. Evidence rated with moderate or high certainty suggested a reduced risk of T1D in relation to longer breastfeeding and later introduction to gluten, cow's milk, and fruit, and an increased risk of IA conferred by high childhood intake of cow's milk. Associations were also seen between T1D and childhood intakes of carbohydrates, sugar, sugar sweetened beverages, protein, meat, nitrite, vitamins A and C, as well as vitamin D supplementation in infancy. However, the certainty of this evidence was graded as low. The small number of identified RCTs has failed to find beneficial effects of dietary interventions.This study supports that diet may play a role in the development of T1D. However, an important conclusion is that although many dietary factors are associated with T1D, the certainty of this evidence is low. There is a clear need for future high quality observational studies, preferable with objectively assessed nutritional information and detailed adjustments for confounders, to further elucidate the influence of diet in the etiology of T1D. We also need clinical trials to evaluate if diet modification may indeed prevent T1D.Alt-text: Unlabelled box1Type 1 diabetes (T1D) is an autoimmune disease characterized by destruction of the insulin producing beta-cells that results in chronic dependence on exogenous insulin Despite decades of research, the role of environmental factors in the etiology of T1D remains unclear 22.1https://www.crd.york.ac.uk/prospero/display_record.php?ID=CRD42020212505. This study adheres to the PRISMA guidelines. Eligible studies were in English, had IA or T1D as outcome; nutrients, foods, or beverages, or nutritional biomarker levels as exposure; cohort, case-cohort, case-control, nested case-control, or randomized controlled trial (RCT) design; and hazard ratios, risk ratios, or odds ratios with 95% confidence intervals (CI), as measures of association. In the present paper, the term relative risk (RR) was used for all measures of association. Congress papers, editorials, interviews, letters, and animal studies were excluded.A systematic review and meta-analysis was conducted after registration in the International Prospective Register for Systematic Reviews (PROSPERO) th, 2020, by librarians at Karolinska Institutet University Library (for complete search strategy see Supplementary Tables 1-3). Relevant studies were retrieved from the reference lists of eligible articles. Title and abstract of all identified studies were screened and articles that seemed likely to fulfill the eligibility criteria regarding the exposure and outcome definitions, as well as the study design, were fully examined. Study screening, data extraction, and risk of bias and certainty of evidence assessment were performed independently by AML and JEL, and disagreements were resolved by consultation with SC.Medline (Ovid), Embase, and Cochrane Library (Wiley) were searched from inception until October 152.2From eligible articles we extracted name of first author, publication year, country, cohort name (when applicable), study design, sample size, number of cases, sex, age at diagnosis, type and quantity of exposure, reference group, outcome, method of exposure and outcome assessment, presence of IA at baseline, risk genotypes, family history of T1D, follow-up time, RR with 95% CI, and included covariates. When multiple estimates were available, the most adjusted was extracted and when only separate estimates of different population strata  were provided, they were pooled using fixed-effects models before inclusion in meta-analysis. Additionally, when several articles used the same data, only one of these was included (Supplementary Table 4). We contacted the corresponding author when vital information was missing and if no response was obtained, the study was excluded.2.3Risk of bias was assessed with the Risk of Bias in Non-randomized Studies of Interventions (ROBINS-I) 2.4Statistical analyses were performed with Stata Statistical Software Release 16 (StataCorp). Random-effects models were used for estimating summary RR and 95% CI of IA and T1D in relation to the highest versus lowest or continuous exposure to each dietary factor, depending on the availability of estimates. The cut-offs of categorical exposures differed across studies and are therefore presented as ranges in the meta-analyses. Thus, the range represents the cut-offs used in the individual studies, e.g., the cut-off ≥2-3 vs <2-3 months means that some of the included studies used the cut-off of two months, and others used three months. Exposure definitions in individual studies are presented in Supplementary Table 6. If the exposure categorizations were too diverse to pool, we present the RR from the meta-analysis based on the largest number of studies, in 2 statistic . To investigate the source of heterogeneity, subgroup analyses based on study design, risk of bias, and genetic susceptibility, were performed when I2 >50% and at least five studies were available. We also performed post hoc sensitivity meta-analyses restricted to prospective data for the outcome T1D; for the outcome IA all studies had a prospective design. Small study effects were assessed for meta-analyses with at least 10 studies with the Egger's test and contour-enhanced funnel plots with critical regions at 1%, 5% and 10% significance levels were used to discriminate whether small study effects were attributable to publication bias.Heterogeneity across studies was assessed with the Cochran's Q test and the percentage of observed variance due to heterogeneity with the I2.5The certainty of evidence for each meta-analysis was assessed through the Grading of Recommendations, Assessment, Development and Evaluations (GRADE) framework, which classifies the certainty of evidence as ‘very low’, ‘low’, ‘moderate’, and ‘high’ 2.6The funders had no role in study design, data collection, analysis, interpretation, or writing of the report.33.1Out of 5,935 articles initially screened, 96 studies with a cohort (n=42) or case-control (n=54) design could be meta-analyzed . Of thes42=43%). A tendency towards an inverse association was noted for later introduction to infant formula, whereas age at introduction to solid food was unrelated to T1D, although with substantial heterogeneity across studies (Longer durations of any or exclusive breastfeeding were inversely associated with T1D . The lar studies . Neither studies .4.1Later introduction to cow's milk (≥2-3 versus <2-3 months) was inversely associated with T1D , whereas4.2For T1D, positive associations were observed in relation to childhood intake of meat, protein, and nitrite, with no heterogeneity between studies . In cont4.32=98%) and IA  in relation to higher childhood intake of omega-3 fatty acids, but the associations were not significant and there was substantial heterogeneity (2=0%), but no data on a potential association with T1D. There was no indication of a reduced risk of T1D in relation to childhood fish intake.The results were compatible with a reduced risk of T1D , with a similar tendency for age at introduction to vegetables  and childhood intake of fruit juice  . Higher 4.52=0%) with a similar tendency for cereal introduction  and, for gluten introduction, possibly explained by genetic susceptibility . Gluten intake in childhood showed a tendency towards a positive association with both T1D and IA but confidence intervals were wide, and heterogeneity substantial (Later introduction to gluten (3-6 versus <3-5 months) was associated with reduced T1D risk  and without any indication of heterogeneity across studies .Vitamin D supplementation during infancy was inversely related to T1D , but theThe risk of T1D and IA was also investigated in relation to omega-6 fatty acid intake during childhood and maternal coffee intake, but no associations were observed  and 3.4.8The presence of publication bias could be assessed for meta-analyses investigating the risk of T1D in relation to any breastfeeding and cow's milk introduction . No publication bias was observed through visual inspection of the funnel plots, however, the Egger's test indicated presence of small study effects for any breastfeeding (p=0.0059). Meta-analyses restricted to prospective studies of T1D (25 out of 80 studies) could be performed for less than half of the dietary factors, including breastfeeding (any and exclusive); age at infant formula, solid food, meat, fruit, vegetable, cereal, and gluten introduction; gluten and fiber intake during childhood; maternal, infant, childhood, and adult vitamin D levels; and omega-3 and omega-6 intake during infancy or childhood . These analyses revealed significantly reduced risks of T1D in relation to later fruit and gluten introduction and higher 25(OH)D levels during adulthood . Only one prospective study assessed the risk of T1D in relation to longer breastfeeding (≥12 versus <12 months) and showed an inverse association .4.9Associations between T1D and breastfeeding (≥6-12 versus <6-12 months) and age at introduction to gluten, were rated with high certainty, and associations with exclusive breastfeeding, age at introduction to cow's milk, and fruit, with moderate certainty . For IA,4.10Studies that were not meta-analyzed include 20 observational studies and six RCTs that either had overlapping study populations or were evaluating different dietary factors (Supplementary Tables 7-9). The observational studies assessed introduction to probiotics, maternal micronutrient intakes, and nutritional biomarker levels during infancy and childhood. The RCTs assessed the effects of weaning to an extensively hydrolyzed versus cow's milk-based formula, introduction to gluten at 12 months versus six months, and nicotinamide supplementation in childhood. No beneficial effects were seen for these factors (Supplementary Table 7).5This meta-analysis included 26 dietary exposures assessed in 96 studies. The results indicate positive associations between T1D or IA and childhood intake of cow's milk, carbohydrates, sugar, sugar-sweetened beverages, protein, meat, and nitrite. In contrast, inverse associations were observed with breastfeeding, later introduction to cow's milk, fruit, and gluten, vitamin D supplementation in infancy, 25(OH)D levels in adulthood, and childhood intake of vitamin A and C. None of the fetal dietary exposures were related to IA or T1D. The certainty of evidence was high for a reduced risk relating to longer breastfeeding and later introduction to gluten; and moderate for an increased risk associated with higher intake of cow's milk, and a reduced risk with exclusive breastfeeding, and later introduction to cow's milk and fruit. For the remaining dietary factors, certainty was low . This isOur results in favor of beneficial effects of breastfeeding are consistent with previous meta-analyses,,13 and aOur study also indicated adverse effects of cow´s milk both in terms of quantities consumed in childhood and early introduction. It has been speculated that the excess risk of T1D associated with earlier introduction to cow´s milk may reflect lack of breastfeeding Carbohydrates, sugar, and sugar-sweetened beverages were associated with increased risk of T1D, although certainty of this evidence was low primarily owing to serious risk of bias in the individual studies. The underlying mechanism could involve effects on body weight, since adiposity is associated with increased risk of T1D Vitamin D is suggested to play a role in the etiology of T1D, primarily due to its involvement in the regulation of the immune system Intake of antioxidants may hypothetically reduce the risk of T1D by inhibiting oxidative stress. Consistent with this hypothesis, we found a reduced risk of T1D in relation to vitamin A and C exposure during childhood, although the certainty of this evidence was very low.Strengths include a broad and systematic literature search according to an a priori defined protocol, and assessment of different types of bias in the individual studies and of the overall quality of evidence for each dietary factor that was meta-analyzed. By performing study selection, data extraction, and risk of bias assessment in duplicate, we reduced the risk of missing important information and introducing errors. Additionally, we distinguished between exposures in utero, infancy, and childhood. The main limitation is that the number of studies was small for most dietary factors. This reduced power and the possibilities to perform sub-group analyses, assess dose-response relationships, and publication bias. Several of the included studies had serious risk of bias, mainly due to confounding, and potential misclassification of exposure caused by self-reported dietary information. The latter is a particular concern in case-control studies with retrospective information. This resulted in reduced certainty of evidence for most associations. Importantly, biomarker information was used in some studies, which reduces this bias. The GRADE tool does not distinguish between prospective and retrospective evidence. However, we performed sensitivity analyses restricted to prospective studies, which substantially reduced the number of eligible studies but revealed similar associations for the dietary factors that could be analyzed. In parallel, RCTs were few and not possible to synthesize since the interventions differed. A main concern regarding confounding was the lack of adjustment for other dietary factors, considering that no food is consumed in isolation. Associations were generally weaker for IA than for T1D. This could reflect that IA is a less specific outcome, identified by the presence of one or several different autoantibodies with varying positivity cut-offs. The assessment of dietary factors varied within meta-analyses both in terms of definition, timing, and categorization. This implies that some studies could not be included in the meta-analyses, e.g., if they had a unique exposure assessment that did not align with that of other studies. Most importantly, it precludes formulation of specific recommendations, e.g., regarding optimal age at introduction of certain foods, quantities of milk consumption, and duration of breastfeeding. Finally, it should be noted that most studies came from Europe and North America and whether the results apply to other populations with different genetic makeup and dietary habits remains to be explored.Our findings indicate that longer breastfeeding and delayed introduction to gluten, cow's milk, and fruits may prevent T1D. Moreover, we find support for the general recommendation of vitamin D supplementation in early life. The potential harm in terms of T1D risk of high cow's milk consumption in childhood may be considered in risk-benefit analyses underlying dietary guidelines. There were several dietary factors previously linked to T1D for which we did not find associations including iron, fish, and omega-3 fatty acids and as noted above, many of the observed associations were of low certainty. This emphasizes the need for future studies to elucidate the role of diet in the etiology of T1D and these studies should preferably use a prospective design, nutritional biomarkers, and careful adjustment for potential confounders. Moreover, the potential interaction of genetic susceptibility with diet needs to be explored. Few studies investigated dietary factors in relation to progression from IA to T1D. This is an important topic for future studies as it may provide information to be used for prevention. In addition, distinguishing between stage 1 (IA with normoglycemia) and stage 2 (IA with dysglycemia) of presymptomatic T1D, when studying progression, may provide further insights in the etiology of T1D In conclusion, this study supports that infant and childhood diet may influence the risk of T1D. The most convincing evidence was seen for beneficial effects of breastfeeding, later introduction to gluten, cow´s milk, and fruit, and lower consumption of cow´s milk during childhood. Other associations were observed, but the certainty of evidence was low. The results emphasize the need for future, high quality studies on the influence of diet on development of T1D.6The study was conceived by SC together with AML and JEL. AML and JEL independently performed the study selection, data extraction, and risk of bias and certainty of evidence assessment and consulted SC in case of disagreement. AML did the statistical analyses and wrote the first draft with input from JEL and SC, who also edited the manuscript. All authors contributed to the interpretation of the findings, had full access to all the data and final responsibility for the decision to submit for publication.The authors declare no competing interests.


Background and Objectives: At present, the association between the long non-coding RNA (lncRNA) metastasis-associated lung adenocarcinoma transcript 1 (MALAT1) polymorphism rs3200401 C > T and cancer risk remain controversial. The aim of this meta-analysis was to assess the association between rs3200401 C > T and cancer susceptibility. Materials and Methods: The databases of PubMed, EMBASE and Web of Science were searched for literature published in English until 1 September 2021. The odd ratios (ORs) and 95% confidence intervals (CIs) were applied to evaluate the strength of association in five genetic models. Heterogeneity was assessed using the Q-test and I2 test. Begg’s funnel plot and Egger’s linear regression test were conducted to assess publication bias. Meta-regression analysis was used to explore potential sources of heterogeneity. Trial sequential analysis (TSA) was performed to validate the reliability of the results. Results: A total of 10 case–control studies involving 6630 cases and 7457 controls were included in this study. The pooled ORs showed no significant association between MALAT1 rs3200401 C > T and cancer risk in five genetic models. Similarly, the association was not found in the subgroups of control source, ethnicity and study quality. In the cancer type subgroup, the results demonstrated that the T allele increased the risk of colorectal cancer (CRC) compared with the C allele. . Conclusion: In the current meta-analysis, we found no significant association between MALAT1 polymorphism rs3200401 C > T and overall cancer risk. However, the rs3200401 C > T may be linked to a higher risk of CRC, which needs more studies to be further confirmed. In recent years, cancer incidence and mortality rates have increased rapidly worldwide. In 2020, there were 19.29 million new cancer cases and 9.96 million cancer deaths worldwide . TherefoMALAT1) is a nuclear-retained lncRNA, more than 8000 nucleotides in length, located on chromosome 11q13 [MALAT1, which has been found abnormally overexpressed in multiple cancer tissues, is involved in cell cycle regulation, and regulates alternative splicing, pre-mRNA splicing and interacts with miRNA [MALAT1 promotes the progression, invasion and metastasis of cancer to a certain extent [MALAT1 enhances the expression of serine-rich arginine splicing factor 1 (SRSF1) and activates the mammalian rapamycin target (mTOR) signaling pathway to promote the formation of gastric cancer (GC) and hepatocellular carcinoma (HCC) [MALAT1 inhibits epithelial–mesenchymal transition and induces apoptosis in tumor cells [Long non-coding RNAs (lncRNA), a type of RNA with a length of more than 200 bp, lack functional open reading frames (ORFs) and protein-coding capabilities . They hame 11q13 . MALAT1,th miRNA . Thus, Mn extent . Recent ma (HCC) . Additioor cells ,16.MALAT1 polymorphism rs3200401 C > T and cancer risk has been investigated in recent years. Peng et al. showed that females with a CT genotype of rs3200401 had a lower risk of breast cancer (BC) [MALAT1 gene is associated with an increased risk of esophageal squamous cell carcinoma (ESCC) [MALAT1 rs3200401 C > T and the risk of cancer.The presence of single nucleotide polymorphisms (SNP) may directly or indirectly influence lncRNA expression levels to regulate the occurrence and development of cancers . SNPs hacer (BC) . Similarcer (BC) . There icer (BC) . Qu et aa (ESCC) . In EuroThis meta-analysis was conducted following the Preferred Reporting Items for Systematic Reviews and Meta-Analysis (PRISMA) Statement.MALAT1” AND “neoplasm OR cancer OR tumor OR neoplastic OR carcinoma OR adenocarcinomas OR malignancy OR malignancies OR neoplasia” AND “single nucleotide polymorphism OR SNP OR variant OR variation OR polymorphism”. We have also carefully screened references of relevant publications to obtain potential studies.A comprehensive search was performed on online databases, including PubMed, EMBASE and Web of Science up to 1 September 2021. The following search terms were used in the databases: “Publications obtained through the search meet the following inclusion criteria: (1) the relationship between SNP rs3200401 C > T and cancer risk was described in case–control studies; (2) odd ratios (ORs) and 95% confidence intervals (CIs) can be estimated via the frequencies of genotypes or alleles; (3) the publications were in English only; (4) the data of the control group satisfied the Hardy–Weinberg equilibrium (HWE). The exclusion criteria for publications are as follows: (1) reviews, letters, case reports; (2) duplicate publications; (3) studies without sufficient or qualified data. In addition, two authors (K. Li and Z. Han) independently checked the relevant articles to assess whether studies met the criteria.p value of HWE and genotyping methods. If the results of two reviewers were inconsistent, consensus was reached through discussion with the third reviewer (J. Wu). The Newcastle–Ottawa quality assessment scale (NOS) was used to assess the quality of included studies [The data of included studies were extracted by two independent reviewers (K. Li and Z. Han) according to the following contents: first author, publication year, region, ethnicity of study population, source of control, cancer type, genotypes of rs3200401 in case and control group,  studies . We rate studies .MALAT1 rs3200401 C > T and cancer risk. Five genetic models  were applied for the analyses. To assess the heterogeneity of all studies, the Q-test and I2 test was performed. If the results showed I2 > 50% or p < 0.05, we would consider the heterogeneity to be significant and use a random-effect model (the DerSimonian–Laird method). If not, a fixed-effect model (Mantel–Haenszel method) would be used. Subgroup analysis of the quality of studies, type of cancer, ethnicity and source of control were performed to explore potential associations. Based on deleting medium-quality studies, sensitivity analysis was performed to assess the stability of the results by observing the alteration after excluding every single study in turn. Publication bias was evaluated using Begg’s funnel plot and Egger’s regression test. Meta-regression analysis was used to identify whether some factors were the source of heterogeneity. Trial sequential analysis (TSA) was carried out to assess the reliability of the results. The TSA parameter was set to a power of 80%, type I error of 5%, relative risk reduction of 15% and control event proportion of an average of each included study. STATA software version 15.1  and TSA software version 0.9.5.10  were applied for statistical analysis.The indictors of ORs and 95%CI were used to assess the relationship between The screening procedure is shown in 2 = 48.8%, p = 0.040). Therefore, the combined data were calculated with a random-effect model in allelic models, and a fixed-effect model was used in heterozygote, homozygote, dominant and recessive models. The results of crude analysis indicated no significant association between rs3200401 C > T and cancer risk in all genetic models  in turn. The results suggest that ethnicity, source of control and quality of study were not statistically confirmed as confounding factors . We performed a trial sequential analysis to reduce the random errors and strengthen the robustness of the association between rs3200401 C > T and CRC risk. As shown in MALAT1 and cancer risk. MALAT1 rs3200401 C > T polymorphism has also been extensively studied in relation to different cancer risks. However, the results of these studies have been inconsistent. We performed a meta-analysis to identify the role of rs3200401 C > T in cancer susceptibility. The crude analysis results showed no association between rs3200401 C > T and cancer risk in the five genetic models. We observed that Asians or Caucasians had no significant effect on the correlation between rs3200401 C > T and cancer susceptibility, suggesting that ethnicity is not a potential confounding factor. However, the current meta-analysis included only two European studies, and more studies are needed to confirm the results. Subgroup analysis by cancer type revealed that rs3200401 C > T increased the risk of CRC in the allelic model.Recently, many studies have focused on the relationship between genetic variation in MALAT1 is enriched in nuclear spots and influences the distribution of serine/arginine (SR) family splicing factors in the region. SR proteins can regulate the alternative splicing (AS) of pre-mRNA and alter the susceptibility of cancers [MALAT1 to SRSF2. The mutation of rs3200401 may lead to the downregulation of SRSF2 phosphorylation and a change in tumor-related gene shearing, thus affecting the occurrence and development of cancer [MALAT1 was widely reported to be overexpressed in HCC patients [MALAT1 expression in HCC needs to be further investigated. The results of the subgroup analysis showed that rs3200401 T allele was significantly associated with increased CRC risk compared with C allele. One study reported a tendency that serum MALAT1 expression levels of the TT genotype and CT + TT genotype were higher than that of the CC genotype in CRC patients [MALAT1 expression levels between rs3200401 C>T and cerebral ischemic stroke [MALAT1 mRNA was overexpressed in CRC tissues according to the Oncomine expression profiling database [MALAT1 expression promotes the development, invasion and metastasis of CRC through multiple pathways and is associated with poor prognosis [ cancers . rs32004f cancer . The assf cancer . Howeverpatients ,34, whicpatients . In our patients . A similc stroke . Li et adatabase . The uprrognosis ,40. Lamprognosis . These srognosis . Despiterognosis . TherefoThis is the first meta-analysis to assess rs3200401 C > T polymorphism associated with cancer risk to our knowledge. The included studies were conducted in recent years, which reduced heterogeneity to some extent. We performed a relatively comprehensive subgroup analysis to explore potential heterogeneity. There were also several limitations in our study. Firstly, it is insufficient because of the case–control studies enrolled in our study. Studies involving CRC are limited, and more studies of larger sample sizes are needed to verify the association between rs3200401 C > T polymorphism and CRC risk Secondly, we only retrieved the studies published in English. This may have caused us to ignore some studies published in other languages. Finally, due to the limited inclusion of studies, studies with low NOS scores were not excluded.MALAT1 polymorphism rs3200401 and cancer risk.In summary, the study indicated that the association between rs3200401 C > T and overall cancer risk was not significant. However, rs3200401 C > T may increase the risk of CRC. Larger sample size studies on a wide range of cancer types are required to perform further verification about the relationship between


Type 2 diabetes mellitus (T2DM) is one of the most common chronic diseases in adults, causing high morbidity and mortality worldwide. In recent years, the prevalence of T2DM has been increasing significantly, and genome-wide association studies (GWAS) have shown that KCNQ1 significantly increases the risk of T2DM.  To find large-scale evidence on whether the KCNQ1rs2237892C⟶T gene polymorphism is associated with T2DM susceptibility.  A comprehensive review of the Chinese and English literature on the association of T2DM with KCNQ1rs2237892 is published by PubMed and Baidu Academic. The included literature was part or all of the studied loci which were evaluated for association with T2DM. Forest plots were made of the included literature to analyze the association of KCNQ1 with polymorphisms of the studied loci, and funnel plots and Egger's test were used to evaluate the publication bias of the selected included literature. P < 0.00001), recessive  genetic model under the full population was observed between KCNQ1rs2237892C⟶T gene polymorphism and T2DM without a significant relationship. In a stratified analysis by race, a meaningful association was found in non-Asian populations under the allelic genetic model, but no association was found in Asian populations.  Ten case-control studies including a total of 7027 cases and 8208 controls met our inclusion criteria. Allele  (OR: 1.19; 95% CI: 0.87,1.62;  This meta-analysis showed no significant association between the rs2237892 polymorphism of the KCNQ1 gene and the risk of T2DM. Type 2 diabetes mellitus (T2DM) is a genetically heterogeneous metabolic disorder characterized by chronic hyperglycemia due to impaired insulin secretion and sensitivity influenced by genetic and environmental factors. Patients with T2DM are often associated with macrovascular disease, diabetic retinopathy, diabetic nephropathy, and diabetic neuropathy . The GloTherefore, we performed this meta-analysis to further demonstrate whether genetic factors play a crucial role in the pathogenesis of T2DM.The advanced search of the literature search library was conducted by using “T2DM, KCNQ1, rs2237892” as the search term in China National Knowledge Infrastructure (CNKI) and Baidu Academic with the following search formula: subject (T2DM) and keyword (KCNQ1) and keyword (rs2237892). English literature was obtained for case-control studies and cohort studies on the association of KCNQ1 gene polymorphism with T2DM. The last search was conducted on June 12, 2021. Inclusion criteria were (i) meeting the diagnostic criteria for diabetes mellitus published by the World Health Organization in 1990 or ADA in 2010; (ii) the study type was a case-control study; (iii) there were year of publication; there were clear regulations on sample content; the study results were informative enough for analyzing whether the differences in genotypes and alleles between the case and control groups were statistically significant; (iv) the control groups all met the H-W genetic equilibrium pattern; (v) sources of information on control and case groups were provided; and (vi) patients were randomly selected, with no special restrictions on age, sex, or family history. The exclusion criteria were (i) lack of sufficient control group; (ii) exclusion of literature review; (iii) exclusion of studies with gestational diabetes as an endpoint; and (iv) insufficient sample size.Two investigators independently performed literature reading and information extraction from the eligible literature based on exclusion and inclusion criteria. When ambiguities were encountered, agreement was eventually reached in whether to extract data from the papers by conferring with the third investigator. For each paper, the following paper information was collected: (i) author's name; (ii) year of publication; (iii) country; (iv) number of included cases and controls; (v) allele and genotype data; and (vi) mean age of included cases and controls. The literature screening process is shown in P < 0.05. If there was heterogeneity between individual studies (I2 > 50%), a random effects model was used to calculate combined effect estimates; otherwise, a fixed effects model was used, and the Z test was used to determine combined OR significance. Potential publication bias was estimated using funnel plots. The degree of asymmetry was statistically assessed using the Egger unweighted regression asymmetry test using STATA 11.0 software.Statistical analysis was completed using RevMan 5 software, and the OR values and their corresponding 95% CIs were used as criteria for data statistics to compare rs2237892 allele distributions. Allelic , recessive (TT vs. CC+CT) genetic models were used, and the significance level was set at We obtained studies on association between diabetes and gene locus polymorphism from Baidu Academic and China Knowledge Network, and some literature had duplicate publications, such as, double submission in Chinese and English, multiple submissions in one manuscript, and overlapping databases. After reading the titles and abstracts of the papers for the first screening, and reading the full text for the second screening, 10 papers were finally included. The required data were recorded by reading the full text to form a dataset for meta-analysis. The number of T2DM patients in the group included in the meta-analysis was 7027, and the number of controls was 8208, with 8 datasets for studies originating from Asia and 2 datasets for studies originating from non-Asia. Information on the first author and year of publication, sample size, mean age of control and case groups, genotype data, and matching criteria for each study are shown in In assessing the effect of the KCNQ1 rs2237892 locus polymorphism on the occurrence of T2DM susceptibility, a total of 10 studies were included in the meta-analysis after a literature data search. The association between T2DM risk and rs2237892 locus polymorphism was assessed using recessive model and allelic model, respectively. Most of the populations were from Asia, so stratified analysis was performed for Asian and non-Asian populations, and the results are shown in Figures 2P = 0.28) and recessive  genetic model was observed between KCNQ1rs2237892C⟶T gene polymorphism and T2DM in the total population. In subgroup analysis stratified by race, individuals carrying the C allele in the Asian subgroup were not associated with T2DM incidence observed under the allelic  genetic model, and recessive  genetic model individuals with CT+TT genotype in the Asian subgroup were not significantly predisposed to increased risk of T2DM. The recessive  genetic model observed no association between CT+TT genotype and T2DM incidence in non-Asian subgroups. However, a significantly increased risk of T2DM patients carrying the C allele was observed in the non-Asian subgroup under the allelic  genetic model.We found significant heterogeneity in the total population and Asian subgroups, and non-Asian populations reflected significant homogeneity see , so a raIn Asian populations, two independent GWASs have identified KCNQ1 as a susceptibility gene for T2DM, but various studies have shown conflicting results.β-cell dysfunction as the main pathophysiological features, and islet β-cell dysfunction is the determinant of DM development. Mutations in any specific protein gene involved in glucose recognition, insulin processing, or secretion can lead to islet β-cell dysfunction. Based on previous studies, it is hypothesized that regulation of potassium channels increases KCNQ1 protein expression on pancreatic β-cells, which reduces insulin secretion and raises blood glucose levels. However, the contribution of KCNQ1 to the pathogenesis of type 2 diabetes is currently unclear [The KCNQ1 gene, a member of the voltage-dependent potassium channel family, is located on chromosome 11 at 11p15.5 and expressed primarily in the heart and less frequently in the pancreas, placenta, lung, liver, kidney, brain, and adipose tissue. Mutations in KCNQ1 cause K+ channel dysfunction, which leads to cardiac long QT syndrome, familial atrial fibrillation, etc. In addition, its mutation has been found to be associated with hearing loss. T2DM development and progression are characterized by insulin resistance and islet  unclear .In this paper, we performed allelic model and recessive gene model meta-analysis of the 10 included studies on the KCNQ1rs2237892 locus and T2DM. Meta-analysis revealed no significant differences in the frequencies of the three genotypes  and two alleles (C and T) of KCNQ1 rs2237892SNP in the case and control groups. The finding is inconsistent with previous studies in which Japanese researcher first found a statistically significant association between the KCNQ1 rs2237892 locus and T2DM onset in an Asian population in a whole gene chain scan study  and did In conclusion, our investigation cannot prove the association of KCNQ1 rs2237892SNP with T2DM in Asian populations, and the observation of the C allele as a risk allele for T2DM in non-Asian populations may be due to small sample sizes , 17, takIn our study, we did not find that the KCNQ1rs2237892 gene polymorphism was associated with T2DM risk, which is inconsistent with the results of other studies. This apparent discrepancy may be mainly due to the different genotype and allele frequencies of KCNQ1rs2237892 in populations with different clinical characteristics, geographic distribution, and ethnic origin. However, the current meta-analysis has some limitations. Large-scale studies on the association of T2DM with the KCNQ1rs2237892 locus are still lacking, and therefore, a larger sample size is needed for further studies in which consideration of all races of subjects may be beneficial to validate the exact results of such observations. These findings may fill a gap in the knowledge system of genetic variation and T2DM-related studies, and therefore, genome-wide association studies in multiple races and geographic regions are necessary.A possible limitation of this study is that the sample size is small, and there may be bias in the results. Our later study will likely increase the sample size and further explore this issue using a stratified cross-sectional design.We concluded that there was no significant association between the rs2237892 polymorphism of the KCNQ1 gene and the risk of T2DM in a meta-analysis of 10 case-control studies from Asia and non-Asia including a total of 7027 cases and 8208 controls.


Background: Recent studies have suggested that proton pump inhibitors (PPIs) and histamine type 2 receptor antagonists (H2RAs) may increase the risk of fracture. We performed a meta-analysis to evaluate the risk of fracture with PPIs and H2RAs use in children and young adults.Methods: PubMed, EMBASE database, Cochrane Library, and Web of Science for relevant articles published before May 2021 were searched. We included all the observational studies reporting on the risk of fracture with acid-suppressive drug (PPIs and H2RAs) use in children and young adults. We calculated pooled risk ratios (RRs) for fracture using random-effects models and conducted subgroup analyses.Results: A total of six studies were included in our analysis. Pooled analysis of PPIs use showed significant risk for fracture , but not significant for PPIs combined with H2RAs use , as well as for H2RAs use alone . Grouping of studies by region showed a significantly increased fracture risk with PPIs use in North America  than in Europe  and Asia . However, there was no significant association between the H2RAs use and the fracture risk in North America . Moreover, PPIs use showed an increased risk of fracture in women , whereas there was no significant association between the PPIs use and the risk of fracture in men .Conclusion: PPIs use alone could increase the risk of fracture in children and young adults, but not for PPIs combined with H2RAs use or H2RAs use alone. Clinicians should exercise caution when prescribing PPIs for patients. Helicobacter pylori infection and gastroesophageal reflux disease (GERD)  . These de (GERD) . AccordiIn adult studies, PPIs use can significantly increase the risk of fracture, particularly fractures of the spine and hip . Recent H2RAs are an alternative medication to treat GERD in children and young adults . One stuThe MOOSE guidelines were adhered for meta-analysis of observational studies and PRISMA guidelines .PubMed, EMBASE database, Cochrane Library, and Web of Science for relevant articles published before May 2021 were searched by two independent reviewers (Li and Xie). We used the following search terms: “proton pump inhibitors” OR “histamine type 2-receptor antagonists” AND “fracture, bone” OR “broken bones” AND “children” OR “young” OR “young adult” OR “infant” OR “adolescent.”Eligible studies were included if they fulfilled the following criteria: (1) cohort or case control studies, (2) reported on the risk of fracture with PPIs or H2RAs use, (3) the reference group were non-PPIs or -H2RAs users, and (4) studies provided adequate data for the risk estimates. The criteria for exclusion were as follows: (1) studies that only reported on bone density changes, (2) duplicated articles, (3) case reports, reviews, meta-analyses, editorials, and letters, (4) insufficient data, and (5) studies conducted in animals or cell lines.The following data were extracted from each study: name of the first author, publication year, study design, country, sample size, age, sex proportion, study period, total participant, duration of follow-up, and effect sizes (risk ratios (RRs), odds ratios (ORs), and hazard ratios (HRs)). The quality of included studies was evaluated using the Newcastle–Ottawa Scale (NOS) , whereinI2 statistic and Q statistic were used to assess the heterogeneity across the included studies. I2 > 50% and p < 0.05 indicated significant heterogeneity across studies. We performed a sensitivity analysis by removing one study per time to test the robustness of the results. The publication bias of the included studies was examined by the Begg funnel plot and Egger test.Analyses were performed using the Stata 12.0 software package. Because the incidence of fracture is sufficiently rare (<5% per year), ORs were considered as approximates of RRs. We extracted HRs, ORs, and RRs from the included studies; otherwise, we calculated unadjusted risk ratio using the raw data. The pooled risk ratio (RR) with 95% confidence intervals (CIs) from HRs and ORs was calculated using a random-effects model. The We initially identified 114 potentially eligible studies after the literature search process, and after removing 31 duplicated ones, 83 studies remained, in which we excluded 57 other studies which were obviously irrelevant based on the titles and abstracts. Also, after full-text assessment, we eventually identified six studies that could be used for our meta-analysis. The literature search process is illustrated in Six studies  were incI2 = 79.3%). In contrast, the H2RAs use did not show a significant association with the fracture risk . Interestingly, PPIs combined with H2RAs use also didn't increase the risk of fracture .As shown in I2 = 35.2%).I2 = 0.0%) and Europe  than that in Asia . There was no significant association between the H2RAs use and the fracture risk in North America ; however, the H2RAs use significantly increased the risk of fracture in Europe . There was also a significant positive association between the combined PPIs and H2RAs use and the fracture risk in North America .When we grouped studies by region, we observed that the risk of fracture in patients with the PPIs use was higher in North America  and low-quality ones . In contrast, we found no association between the H2RAs use and the fracture risk in both the high-quality studies  and low-quality ones . Moreover, H2RAs combined with PPIs use has a significant association with the fracture risk in the high-quality studies , but not in the low-quality ones .Grouping of studies by NOS score showed a significant positive association between the PPIs use and the fracture risk in both the high-quality studies  and those with less than 100,000 participants included .When we grouped studies by sample size, we found a significant association between the PPIs use and the fracture risk in both the studies with more than 100,000 participants included , and other fracture .In the subgroup meta-analyses by fracture outcome, the PPIs use increased the risk of low limb fracture , whereas there was no significant association between the PPIs use and the risk of fracture in men .The subgroup analyses by sex showed a significant association between the PPIs use and the risk of fracture in women .In this meta-analysis concerning the six selected observational studies, we found an increased risk of fracture with the PPIs use alone or the PPIs combined with H2RAs use in children and young adults, but not with the H2RAs use alone, which is consistent with the meta-analysis results of the acid-suppressing medication and fracture risk in adults . BesidesPPIs use is becoming increasingly common, and there has been an observed increase in the risk of gastrointestinal and respiratory tract infection , hypomagOur meta-analysis has several strengths. First, there were several previous meta-analyses on the fracture risk with PPIs or H2RAs use in the elderly. However, the current meta-analysis is the first to review the PPIs use and fracture risk in children and young adults. Second, it examined the associations stratified by the type of agent (PPIs or H2RAs), the study design, region, NOS score, outcome, and sample size, as well as sex. However, our meta-analysis also has several limitations. First, there was no available RCT on the issue, so our study consisted entirely of observational studies, which are also susceptible to various biases. Second, the number of eligible studies selected in our meta-analysis was relatively small. Third, some potential studies might have been missed with the language of publications restricted to English only.In summary, this meta-analysis of observational studies showed that PPIs use alone could increase the risk of fracture in children and young adults, but not for PPIs combined with H2RAs use or H2RAs use alone. Clinicians should carefully assess the risk of fracture before prescribing PPIs for patients. Further studies are required to confirm our findings reported hereby due to the inevitable limitations of this meta-analysis.


The relationship between ERCC gene polymorphism and osteosarcoma risk / overall survival of osteosarcoma is still conflicting, and this meta-analysis was performed to assess these associations.The association studies were identified from PubMed, and eligible reports were included and calculated using meta-analysis method.Four studies were included for the association of ERCC gene polymorphism with osteosarcoma risk, and nine studies were recruited into this meta-analysis for the relationship between ERCC gene polymorphism and overall survival of osteosarcoma. The meta-analysis indicated that ERCC1 rs3212986 (8092 C>A) gene polymorphism, ERCC1 rs11615 (19007 T>C) gene polymorphism, ERCC2 rs1799793 (A>G) gene polymorphism, ERCC2 rs13181 (Lys751Gln) gene polymorphism were not associated with osteosarcoma risk. ERCC1 rs2298881 (C>A) gene polymorphism, ERCC1 rs3212986 (8092 C>A) gene polymorphism, ERCC1 rs11615 (19007 T>C) gene polymorphism, ERCC2 rs1799793 (Asp312Asn) gene polymorphism were not associated with overall survival of osteosarcoma. Interestingly, ERCC2 rs13181 A allele and GG genotype were associated with overall survival of osteosarcoma, but AA genotype not .ERCC2 rs13181 A allele and GG genotype were associated with overall survival of osteosarcoma. Human osteosarcoma, one of the most familiar forms of the primary malignant tumor to adolescents and adults, is a genetically heterogeneous bone malignancy with poor prognosis despite the employment of aggressive chemotherapy regimens Within DNA repair genes, there lie a number of single nucleotide polymorphisms which may impair protein function and attenuate DNA repair capability, resulting in genomic instability and individual predisposition to malignancies The search was conducted in the databases of PubMed on May 1, 2019, and the relevant investigation were included. The retrieval strategy of “(ERCC2 OR ERCC3 OR ERCC1 OR excision repair cross-complementing) AND (osteosarcoma OR bone tumour OR bone cancer OR bone carcinoma) AND polymorphism” was entered into the PubMed database.Inclusion criteria: (1) The outcome must be osteosarcoma or overall survival of osteosarcoma; (2) The study included two comparison groups (case group vs control group); (3) report should give the data of ERCC genotype distribution.Exclusion criteria: (1) Case reports, editorials and review articles; (2) Preliminary result not on ERCC gene polymorphism or osteosarcoma / overall survival of osteosarcoma; (3) Investigating the role ERCC gene expression to overall survival of osteosarcoma.The following information from each recruited investigation was extracted by two investigators independently (Guanliang Wang and Jianping Li): first author's surname, year of publication, ethnicity, control source of the control group and the number of cases and controls for ERCC genotypes. Frequencies of allele of ERCC were calculated for case group and control group. Quality assessment was assessed by Newcastle-Ottawa Scale (NOS) score, and it was regarded as a high quality (or low-bias risk) study when total stars achieved six to nine.Cochrane Review Manager Version 5  was used in this meta-analysis to count the extracted data from each report. The pooled statistic was counted using the fixed effects model. However, a random effects model was conducted when the P value of heterogeneity test was less than 0.1. Results were expressed using odds ratios (OR) for dichotomous data. 95% confidence intervals (CI) were also calculated. P < 0.05 was required for the pooled OR to be statistically significant, and I2 was used to test the heterogeneity among the included studies.The database of Pubmed was searched for this meta-analysis, and 11 studies were eligible and included for this meta-analysis, and the recruited flowchart is shown in Two studies Two studies Two studies Three studies Two studies Four studies Eight studies Seven studies Seven studies This meta-analysis was performed to detect the relationship between ERCC gene polymorphism and osteosarcomas risk, and the relationship between ERCC gene polymorphism and overall survival of osteosarcoma. We found that ERCC1 rs3212986 (8092 C>A) gene polymorphism, ERCC1 rs11615 (19007 T>C) gene polymorphism, ERCC2 rs1799793 (A>G) gene polymorphism, ERCC2 rs13181 (Lys751Gln) gene polymorphism were not associated with osteosarcoma risk. ERCC1 rs2298881 (C>A) gene polymorphism, ERCC1 rs3212986 (8092 C>A) gene polymorphism, ERCC1 rs11615 (19007 T>C) gene polymorphism, ERCC2 rs1799793 (Asp312Asn) gene polymorphism were not associated with overall survival of osteosarcoma. Interestingly, ERCC2 rs13181 A allele and GG genotype were associated with overall survival of osteosarcoma, but AA genotype not.In previous, Li et al ERCC2 rs13181 A allele and GG genotype were associated with overall survival of osteosarcoma. However, more association investigations are required to confirm these associations.


Purpose: To explore the associations between refractive errors and multiple eye health outcomes.Methods: This is an umbrella review based on systematic reviews with meta-analyses. In our study, refractive errors included myopia, hyperopia, astigmatism, and anisometropia. We reconducted the meta-analyses whose primary data were available in sufficient detail by random effect model. Heterogeneity was assessed by I2. The main outcomes included myopic macular degeneration (MMD), retinal detachment (RD), cataract, open-angle glaucoma (OAG), strabismus, age-related macular degeneration (AMD), and diabetic retinopathy (DR).Results: Myopia was associated with increased risk of MMD , RD , nuclear cataract , posterior subcapsular (PSC) cataract , OAG , exotropia , but decreased risk of DR , and early AMD . From mild-to-high myopia, the association strengthened for MMD, RD, nuclear cataract, PSC cataract, OAG, and DR. Hyperopia was associated with an increased risk of early AMD  and esotropia . Astigmatism and anisometropia were associated with increased risk of both exotropia and esotropia.Conclusions: Myopia, especially high myopia, demonstrated the highest risk for eye health outcomes, such as MMD, RD, OAG, nuclear and PSC cataracts, and exotropia. However, myopia was associated with a lower risk of early AMD and DR. Individuals with hyperopia are more likely to suffer early AMD and esotropia. Astigmatism and anisometropia predispose to strabismus. A lot of research studies on the mechanism of the associations are needed.Systematic Review Registration:https://www.crd.york.ac.uk/PROSPERO/display_record.php?RecordID=239744; identifier: 239744 In refractive error, parallel light cannot focus on the fovea to form a clear image after passing through the ocular refractive system; refractive errors include myopia, hyperopia, astigmatism, anisometropia, and presbyopia . In 2010High-quality evidence is needed to demonstrate the associations between refractive errors and eye health. In the past, systematic reviews and meta-analyses were widely considered the highest level of evidence. However, with numerous systematic reviews and meta-analyses available, it is difficult to distinguish priority in most cases. Thus, it may be very useful to perform systematic reviews of systematic reviews, i.e., umbrella reviews . UmbrellWe registered this umbrella review on PROSPERO (CRD42021239744). We reviewed refractive errors and multiple eye health outcomes by systematically searching for meta-analyses. In this review, refractive errors included myopia, hyperopia, astigmatism, and anisometropia. Presbyopia was excluded because we considered that presbyopia is a physiological lens-aging process. Specifically, we excluded systematic reviews without meta-analyses.Two researchers (YW and YT) independently searched PubMed, Embase, the Cochrane Library, China National Knowledge Internet, China Biology Medicine, and the WANFANG database from inception to February 21, 2021, for meta-analyses of observational or interventional studies that analyzed the association between refractive errors and eye health outcomes. We used the following medical subject heading (MeSH) terms and keywords in the search:  and . Two researchers (YW and YT) independently screened the titles and abstracts and then reviewed the full text of the selected articles for eligibility. When differences occurred, a third researcher (CH) was consulted. We also conducted a manual search of the references cited in eligible articles.An article was eligible if it (1) provided odds ratios, risk ratios, hazard ratios, and weighted mean differences, or raw data that could be used to calculate the relative risks or mean difference; (2) was a systematic review with a meta-analysis of observational population-based  and interventional studies ; and (3) investigated the association between refractive error and eye health outcomes. There were no language limitations. If an article contained several meta-analyses, all meta-analyses were included separately. For the same outcome compared in the same group  by multiple articles, we chose only one meta-analysis for each exposure to avoid the inclusion of duplicate studies. The following criteria were used: when the primary articles did not overlap or partially overlapped, we chose the meta-analysis with the largest number of studies, and when the primary studies completely overlapped, we selected the review with the highest quality.I2) and measures of publication bias.Authors YW and YT independently extracted data from eligible systematic reviews and meta-analyses. A third investigator (CH) resolved any difference in the extracted data between the two researchers. For each meta-analysis, the following details were collected: first author, journal, year of publication, populations, number of studies, study design, funding sources, outcome(s) of interest, type(s) of refractive error, relative risks or mean difference and corresponding CIs, the measure of association, numbers of events and non-events, follow-up years (cohort studies), and type of effect model (random or fixed). We also extracted estimates of the proportion of variance .We did not review the original studies of the published systematic reviews and meta-analyses. When the primary data were available in sufficient detail by the systematic reviews and meta-analyses, we reanalyzed each meta-analysis of the extracted data . A randoTwo authors (YT and JZ) independently evaluated the eligible meta-analyses. Disagreements were resolved by consensus. We assessed the methodological quality of the meta-analyses through A Measurement Tool to Assess Systematic Reviews (AMSTAR), which has been proven to be reliable and valid for systematic reviews based on both interventional and observational studies , 12. TheMyopia was the most commonly studied refractive error, followed by hyperopia. Only one study investigated the association between astigmatism or anisometropia and eye health outcomes . The stuMyopic macular degeneration was defined slightly differently across studies see . CompareMyopia was associated with a significantly higher risk of retinal detachment (RD)  . The annThe most frequent definitions of open-angle glaucoma (OAG) were based on glaucomatous visual field loss and optic disc abnormality , 24, 25.According to a summary meta-analysis that combined cohort, case-control, and cross-sectional studies, myopia was significantly associated with a higher rate of nuclear and posterior subcapsular (PSC) cataract . HoweverFrom mild myopia to high myopia, there was an increasing risk of nuclear or PSC cataract . HoweverPeople with myopia or hyperopia had a higher but non-significant risk of having an epiretinal membrane .The diagnosis and classification of AMD were based on the Wisconsin Grading System or the International AMD Classification , 19. MyoPer-diopter increase in SE toward hyperopia was significantly associated with a higher risk of AMD, which was consistent in both prevalent and incident AMD . LikewisThere were consistently associations between myopia and DR, such as any , mild , moderate , and high  myopia. The protective effect increased from mild-to-high myopia, although there was no significance for the association between any/high myopia and DR. The associations between different types of myopia and VTDR were not significant or consistent , 3.Per-diopter increase in SE toward hyperopia was significantly associated with a higher risk of DR . There was also a beneficial but non-significant association between per-diopter increase and VTDR. Furthermore, a per-millimeter increase in axial length was associated with a lower risk of DR and VTDR .Only one study meta-analyzed the association between refractive error and strabismus . All fouI2 <50%. For the 26 meta-analyses that we were unable to reanalyze, 18 had an I2 <50%. Additionally, 18 meta-analyses used the random-effects model.We reanalyzed 58 meta-analyses using a random-effect model. Thirty-nine of the reanalyzed meta-analyses had an P = 0.089) (P = 0.02)   and earl = 0.02) . For metP-values , 18, 19.outcomes , 23, 24.n = 45) were rated as “very low,” 22.4% (n = 15) as “low,” and 10.4% (n = 7) as “moderate.” The median AMSTAR score was 7 across all included studies (range 2–8). The most frequent item downgrading the AMSTAR score was “Was a list of studies (included and excluded) provided?” (14 out of 15), followed by “Was an a priori design provided?” and “Was the status of the publication  used as an inclusion criterion?” (13 out of 15). Because refractive errors, especially myopia, are a global problem, studies have focused on their influence on eye health outcomes and vision impairment , 4, 13. Myopia was strongly associated with a higher risk of MMD, and this association was increased with myopia progression. Studies have reported that the prevalence of MMD ranges from 1.2 to 21.7% in mild myopia, 7.1 to 49.1% in moderate myopia, and 61.3 to 76.4% in high myopia –32. ThesOur meta-analysis demonstrated a higher risk of RD in the myopia group, and the risk was increased for those with more severe myopia. Furthermore, patients with high myopia also had a 2.53-fold risk of re-RD. Han et al. reported a significantly negative association between a per-diopter increase in SE and RD  . For eacThe prevalence of OAG was higher in the myopia group than in the emmetropia group. The risk of OAG was higher in the more severe myopia group. Hollands et al. reported that from a myopia population with diopter ≤ −0.5 D to ≤ −8 D, the prevalence of OAG increased from 3.5 to 10% non-linearly . The mecMyopia was associated with prevalent nuclear and PSC cataracts but not with cortical cataracts. With myopia progression, the risk of nuclear and PSC cataracts increased. Nuclear cataract results from an aging process that is associated with lens fiber accumulation and abnormal lens protein aggregation , 50, whiThe risk of prevalent AMD was significantly lower in the myopia group and higher in the hyperopia group. This association primarily existed in early AMD but not in late AMD. The association between incident AMD and refractive error was not significant. Dose-response analyses also indicated that myopia progression  was associated with a lower risk of early AMD but not for late AMD. However, our re-analysis showed that hyperopia was associated with a lower risk of late AMD unlike noted in a previous meta-analysis . TherefoP < 0.001) and proliferative DR   . Similar< 0.001) . These f< 0.001) . Retinal< 0.001) –73, whic< 0.001) . Therefo< 0.001) –77. As m< 0.001) , 62, whi< 0.001)  reduce r< 0.001) .Our findings showed that myopia was a risk factor for exotropia, while hyperopia increased the risk of esotropia, and astigmatism and anisometropia were associated with a higher risk of concomitant strabismus. A possible mechanism underlying these associations is convergence stimulation. For myopia, the less accommodative effort is required for clear images because of a larger mean lag of accommodation . These fHowever, we should notice that the possible role of misclassification in these analyses. For OAG, it is difficult to clearly distinguish the fundus changes of myopia. Posterior scleral staphyloma and increased stretching force in eyes with longer axial length may increase the C/D ratio and reduce the retinal nerve fiber layer thickness. Besides fundus changes, abnormal OCT RNFL and visual field defects due to peripapillary atrophy and choroidal thinning can also occur in myopes , making Our study is the first umbrella review to systematically summarize the association between refractive errors and eye health. We extracted the data in our study from meta-analyses alone. Our study also provides dose-response analyses, which could help the investigator better assess the relationship between refractive error and different eye health outcomes.There are some limitations to our study. First, except for one RCT (the intervention was not for myopia) , all theThis umbrella review comprehensively analyzed the association between refractive errors and various eye health outcomes. Myopia had harmful associations with most eye health outcomes, such as MMD, RD, OAG, nuclear and PSC cataracts, and exotropia. These associations were more obvious in eyes with high myopia. A decreased risk of DR or early AMD was associated with myopia. The association between myopia and VTDR or late AMD remains uncertain but warrants further investigations. Hyperopia was a risk factor for early AMD and esotropia but was associated with a lower risk of late AMD. Astigmatism and anisometropia showed harmful associations with both exotropia and esotropia. Our results provide higher-level evidence for investigators to understand refractive errors. With the bidirectional effects of refractive errors, especially myopia, the optimal refractive error status should be reconsidered and investigated. Finally, the findings of this umbrella review can be of great clinical interest. The risk of myopia is usually remembered by most clinicians, however, besides the risks of angle closure in hyperopes, the increased risk of AMD in hyperopes should be noted and kept in mind in a routine clinical setting.The raw data supporting the conclusions of this article will be made available by the authors, without undue reservation.Y-hW: conceptualization, methodology, software, formal analysis, resources, data curation, writing-original draft, and visualization. CH: conceptualization, validation, resources, investigation, writing—review, and editing. Y-lT: validation, formal analysis, investigation, writing—review, and editing. JZ: investigation, writing—review, and editing. X-mL: conceptualization, validation, writing—review and editing, supervision, project administration, and funding acquisition. All authors contributed to the article and approved the submitted version.This study was supported by the National Science Foundation of Beijing Municipality [grant number: 7202229].The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


The HRs were 0.93  and 1.26  in Western and Asian countries, respectively. Dose–response analysis indicated a J shaped association, with the lowest risk at 50% carbohydrate intake  and with risk increasing significantly at 70% carbohydrate intake . There was no association between low carbohydrate diet score and the risk of T2D . Carbohydrate intake within the recommended 45–65% of calorie intake was not associated with an increased risk of T2D. Carbohydrate intake more than 70% calorie intake might be associated with a higher risk.We did this study to clarify the association between carbohydrate intake and the risk of type 2 diabetes (T2D) and potential effect modification by geographical location. PubMed, Scopus and Web of Science were searched to find prospective cohort studies of dietary carbohydrate intake and T2D risk. A random-effects dose–response meta-analysis was performed to calculate the summary hazard ratios (HRs) and 95%CIs. The quality of cohort studies and the certainty of evidence was rated using the Newcastle–Ottawa Scale and GRADE tool, respectively. Eighteen prospective cohort studies with 29,229 cases among 607,882 participants were included. Thirteen studies were rated to have high quality, and five as moderate quality. The HR for the highest compared with the lowest category of carbohydrate intake was 1.02 (95%CI: 0.91, 1.15; I It is estimated that at least 514 million people are affected by T2D all over the world1. The Middle East and North Africa is a region with the highest prevalence of T2D across the globe3. T2D is a chronic progressive disease, and therefore, lifestyle modifications such as dietary interventions, physical activity, and weight reduction are the core part of first-line interventions for the prevention of T2D6.Type 2 diabetes (T2D) is a serious non-communicable chronic disease described by impaired insulin action or secretion or impaired response of body cells to insulin, followed by the endocrine pancreas' incapability to compensate for this weakened response7. Of note, dietary carbohydrates have received specific attention because of their effect on blood glucose level8. Dietary carbohydrates are the major dietary energy source10 and have the greatest impact on postprandial blood glucose levels11. Studies have shown that glycemic properties of the diet including glycemic index and load might be associated with the risk of developing T2D and other chronic diseases11.Dietary factors might have a role in the development of T2D13. However, almost all studies included in the published meta-analyses have been conducted in Western countries, where the intake of carbohydrates was lower than that of Asian countries15. A recent publication from the PURE study in 21 countries across the world indicated that higher rice consumption was associated with a greater risk of developing T2D, with the strongest association in South Asia and a modest, nonsignificant association in other regions16.With regard to T2D, three meta-analyses of cohort studies have been undertaken of the association between dietary carbohydrate and the risk of T2D, but the results have been inconsistent19. In the current study, we therefore aimed to update the evidence from prospective cohort studies of the association between dietary carbohydrate intake and low-carbohydrate diet score (LCDS) with the risk of T2D in the general population. Our secondary outcome was to assess this association separately in Asian and Western countries.A number of population-based prospective cohort studies in Asian countries have recently been published that reported significant positive associations between dietary carbohydrates and the risk of T2D20. The protocol of the study was registered at Open Science Framework .The Meta-Analysis of Observational Studies in Epidemiology (MOOSE) guidelines have been used for reporting this meta-analysisWe performed a comprehensive systematic search on all literature issued earlier than April 2021 in online databases including PubMed/Medline, Scopus and ISI Web of Science. We did not exert any limitation in term of language or time of publication. We used search terms relevant to type 2 diabetes, carbohydrate, and study design to find potential eligible cohort studies  published prospective cohort studies conducted in the general population; (2) reported carbohydrate consumption (as either g/d or percentage energy) and LCDS as exposure; (3) considered T2D incidence as the outcomes of interest; (4) provided estimates of the effect size in the form of relative risk, hazard ratio (HR) or rate ratio with corresponding 95% confidence intervals (CIs) for ≥ 2 quantitative categories of carbohydrate consumption or LCDS; and (5) provided the numbers of cases and non-cases or person-years in each category of dietary carbohydrate or LCDS. Studies that reported continuous estimation from the associations were also included. For duplicate publications form the same cohort, the one with the greater number of cases was included in our meta-analysis. We excluded letters, comments, reviews and meta-analyses, and ecologic studies. We also did not include studies that were performed on children or adolescences or those that were conducted among patients with type one diabetes. All outcomes were classified based on the World Health Organization’s international classification of disease criteria.http://plotdigitizer.sourceforge.net/) to extract numerical estimates from graphs.Data extraction process was executed by two reviewers in duplicate (FH and AJ), and any divergences were resolved by consultation the principal investigator (SS-B). We extracted the following information from the publications identified: name of the first author, publication year, country, age, sex, study participants, number of cases, duration of follow-up, method of assessment of carbohydrate consumption and LCDS, the fully-adjusted estimates and their 95%CI and list of potential confounders entered into the multivariable statistical model. Gender-specific estimates were combined a by fixed-effects model to include each cohort once in the main analysis. We used web plot digitizer , smoking status, alcohol drinking, and energy and fiber intakes. 27 and Orsini28 for dose–response meta-analysis. We calculated the HRs for a 10% increment in carbohydrate intake or a 10-point increment in LCDS in each study. Study-specific HRs were combined by a random-effects model. For this purpose, each cohort study must report the number of cases and person-years and median or range of dietary carbohydrate or LCDS across categories of exposures. For studies that reported dietary carbohydrate as g/d, we converted g/d to percentage calorie from carbohydrate by using the average daily energy intake of the study participants. For studies that reported the results per unit increment in dietary carbohydrate , we first converted g/d to percentage energy from carbohydrate and then translated it to a 10% increment in energy intake from carbohydrate. For studies that used different units , we calculated the logarithm of the HR and its 95%CI, multiplied by the corresponding unit, and then exponentiated the results. For studies that reported carbohydrate intake as a range in each category, we used the midpoint of lower bounds as a proxy of the median. The widths of the open-ended categories were considered equal to the closest categories.We used the method introduced by Greenland29. This method estimates the study-specific slope lines and combines them to obtain an overall average slope in a single stage. We included all studies in the main analysis. However, due to substantial difference in carbohydrate consumption in Asian and Western countries, we performed separate nonlinear dose–response analyses in Asian and Western countries. Statistical analyses were conducted using STATA software, version 15.0. P < 0.05 was considered statistically significant.Finally, we performed a one-stage weighted mixed-effects meta-analysis to model dose–response associations30. Accordingly, studies with 1–3, 4–6, and 7–9 points were rated as poor, fair, and high quality, respectively. The certainty in the estimates was rated by the Grading of Recommendations Assessment, Development, and Evaluation (GRADE) approach. GRADE tool is a metric to assess the certainty of the evidence31. This tool grades observational studies as low with downgrades for study limitations, inconsistency, indirectness, imprecision, and publication bias, and upgrades for large effect size, dose–response gradient, and attenuation by plausible confounding.The quality of the original studies included in the present meta-analysis was evaluated using a 9-point Newcastle–Ottawa Scale by two independent investigators (FH and AJ)37 and one study with insufficient data38. Finally, 18 prospective cohort studies were included53 , smoking status (n = 17), energy intake (n = 17), BMI (n = 16), and alcohol consumption (n = 15). Only a few studies included in this meta-analysis did not adjust for energy intake42 and BMI53. Ten studies did not adjust for fiber intake53. Looking at the variation of NOS score, 13 studies out of 18 studies were rated high quality (NOS score of ≥ 7)51, and the others were rated to have moderate quality (NOS score of ≤ 7)53 .Thirteen prospective cohort studies investigated the association between intake of carbohydrates from diet and T2D2 = 6%, n = 6; Table The association did not reach statistical significance by the stepwise exclusion of each primary study at a time (HR range: 0.99 to 1.05). In the subgroup analyses, there was no significant association across subgroups except for studies conducted in Asia  and higher risk as carbohydrate intake increased. The HRs for 60%, 70%, and 80% calorie intake from carbohydrate were, respectively, 1.01 , 1.18 , and 1.41 .All studies but onenonlinearity = 0.43, Pdose-response = 0.01.; n = 7, Fig. Restricting dose–response analyses to studies from Western countries only indicated that the risk of T2D did not change remarkably with increasing carbohydrate intake from 37 to 60% of total calorie , followed by a sharp and linear increment in risk  than in Western countries We updated previous meta-analyses and included the most recent studies, especially those conducted in Asia. We included new Asian articles and looked at them separately by subgrouping them because of the difference in their diet. Here we showed that higher carbohydrate intake more than the recommended daily intake of 45–65% was strongly associated with the risk of T2D. We applied a newly-developed one-stage linear mixed effects meta-analysis that creates more efficient and flexible plots than the conventional two-stage model.Some limitations should be noted in the context of our findings. Due to the observational nature of the studies included, our resulting associations cannot establish causality. According to the GRADE, the certainty of the evidence was rated low for dietary carbohydrate and very low for LCDS. In addition, we had insufficient data for the analysis of LCDS. We used total carbohydrate intake as exposure which represents a large diverse group of foods such as whole and refined grains. The potential difference in foods constituting total carbohydrate intake in Asian and Western countries might confound the association between total carbohydrate intake and T2D.The results of this updated meta-analysis of 18 cohort studies  showed that carbohydrate intake within the recommended dietary intake of 45% to 65% of total calorie was not associated with a higher risk of T2D and even was associated with a modest lower risk at 50% carbohydrate intake. Carbohydrate intake more than 70% of total calorie, as found in Asian countries, was associated with substantial higher risk of T2D. However, these findings were obtained from observational studies and thus, could not prove causality. More research, especially in Asian countries, is needed to investigate the association between carbohydrate intakes higher than recommended dietary intake with the risk of T2D.Supplementary Information.