The overall effect was a statistically significant negative correlation in the random effects model . For SMD, the results showed a lower heterogeneity with a negative SMD that did not reach statistical significance under the fixed effects model . Regarding motor function, heterogeneity (I2 = 75%) was also observed in the correlation approach with a pooled r (random effect model) = −0.150; 95%CI: −0.219 to −0.079; p < 0.001. Moderate heterogeneity was observed according to the SMD approach (I2 = 52.28%), with a pooled SMD = −0.136; 95%CI: −0.188 to−0.084; p < 0.001, indicating worse motor function in those exposed. Conclusions: Correlation approach results support a negative effect on cognitive and motor functions . Regarding the SMD approach, results also support a worse cognitive and motor functions in those exposed, although only for motor function statistical significance was obtained.Background/Objective: Whether environmental exposure to Manganese (Mn) in adults is associated with poorer results in cognitive and motor function is unclear. We aimed to determine these associations through a meta-analysis of published studies. Methods: A systematic review was conducted to identify epidemiological studies on a population ≥18 years old exposed to environmental airborne Mn, and in which results on specific tests to evaluate cognitive or motor functions were reported. We consulted Medline through PubMed, Web of Science and SCOPUS databases. We also performed a manual search within the list of bibliographic references of the retrieved studies and systematic reviews. To weight Mn effects, a random effects versus fixed effect model was chosen after studying the heterogeneity of each outcome. Results. Eighteen studies met the inclusion criteria. Among them, eleven studies reported data susceptible for meta-analysis through a pooled correlation or a standardized means difference (SMD) approach between exposed and non-exposed groups. Regarding cognitive function, the results of the studies showed heterogeneity among them (I Manganese (Mn) is a trace element, and therefore essential for our normal physiological function , being iWith regard to environmental exposure to high levels of Mn, an increasing number of epidemiological studies have been published in recent decades, both in children and in adult populations, focusing mainly on neurological effects on cognitive and motor functions . With reTherefore, the objective of this study is to analyze, through a meta-analysis of published primary studies, the impact of environmental Mn exposure on cognitive and motor functions in adults.A bibliographic search was conducted to identify original epidemiological studies carried out in adults (≥18 years), written in English or Spanish, in which environmental air Mn exposure and/or biomarkers of environmental exposure were assessed, and with at least one determination in cognitive and/or motor functions. We consulted different international bibliographic databases: “Medline through PubMed”, “Web of Science (WOS)” and “Scopus”. We identified all relevant primary studies (published and under publication) until 11th November 2020, by using the strategy: [(MANGANESE OR MN) AND (NEURO* OR COGNIT*) AND ENVIRONMENTAL], using free text and without applying any limitation in the search strategy in PubMed or WOS, and restricting to Title & Abstract in Scopus. We also performed a manual search within the bibliographic references lists of the retrieved studies and systematic reviews. Overall, we found 3310 studies from Medline, 4376 from Web of Science and 1171 from Scopus. Data generation was performed independently by 2 authors and quality  was assessed by using the quality assessment tool for observational cohort and cross-sectional studies from the National Heart, Lung, and Blood Institute (NHLBI), in which 14 criteria are measured to determine an overall quality rating of ‘good’, ‘fair’ or ‘poor’ for each study [s) correlations were available. Pooled estimates of the correlation coefficients were calculated by transforming the correlation coefficients into Fisher’s z values. The resulting values were weighted with the inverse of the variance of the correlation coefficients [A meta-analysis was conducted first restricted to studies for which Pearson’s (r) or Spearman’s rho , with its 95% confidence interval (95%CI), was chosen as a summary measure of the effect to allow us to combine data for the cognitive and motor functions separately. This strategy, which is consistent with the approach taken in other reviews ,19,20, i2 statistic, which describe the percentage of total variation across studies that is attributable to statistical heterogeneity rather than to chance. I2 values of 25, 50 and 75% corresponded to low, moderate and high between-study statistical heterogeneity. A p value < 0.10 was set as the cut-off point for a statistically significant heterogeneity in the chi squared test for heterogeneity [To weight the Mn effects, a random effects versus fixed effect model was chosen after studying the heterogeneity for each outcome. Statistical heterogeneity was assessed through the Cochran’s Q-test and Iogeneity . We usedogeneity . The resWe sought evidence of publication bias by using the funnel plot method and Egger’s regression asymmetry test ,25. In a® “used by Bowler et al., (2016) to determine the Tremor Harmonic index (HI) [Trail Making and Grooved Pegboard Test are based on the time (seconds) required to complete the tests, so the more seconds, the worse function. In the Simple Visual Reaction Time test, the mean is based on milliseconds. Higher scores in the “Unified Parkinson’s Disease Rating Scale (UPDRS)”, Eurythmokinesimetry (EKM) and Postural Balance Testing also denote worse function. In the “Visual Attention Computerized Test, third version (TAVIS-3), both selective and sustained”, the scores are: reaction time, omission errors and commission errors; so longer execution time and greater errors also indicate worse function. This is also the case for the “Coordination Ability Test System (CATSYS) Tremor Pendex (HI) , since adex (HI) . For theAccording to the selection criteria of our systematic-review, 18 original articles were found ,44,45,46Only four articles showed cognitive function data subsidiary to be analyzed through a correlation coefficient in the meta-analysis ,42,43,44s = 0.01) corresponding to the “Trail Making Test A, T-score” [p < 0.001, I2 = 76.49%, Tau = 0.027). The overall effect was that of a statistically significant negative correlation in the random effects model, pooled r: −0.165; 95%CI (−0.214 to −0.116), p < 0.001  out of the 11 different tests meta-analyzed did not correlate negatively. Depending on the domain of cognitive function analyzed, negative correlations were obtained in all domains, reaching statistical significance in seven of the nine domains studied. Depending on the type of exposure to Mn, all exposure matrices were equally associated with negative correlations See .Only three articles showed cognitive function data subsidiary to be analyzed through a SMD between groups in the meta-analysis ,40,41. Tp = 0.425, I2 = 2.62%, Tau = 0.02). The overall effect was a negative SMD that did not reach statistical significance: SMD under the fixed effects model: −0.052; 95%CI (−0.108 to 0.004); p = 0.068 . We obtained similar results regarding the standardized differences of medians: SMD under the fixed effects model: −0.041; 95%CI (−0.097 to 0.015); p = 0.154. SMD under the random effects model: −0.038; 95%CI (−0.102 to 0.025); p = 0.239 , except for intelligence, based on a single Viana determination (SMD = 0.438), and attention and working memory (SMD under the fixed effects model = 0.02). In terms of the type of Mn exposure evaluated, each study showed a different Mn evaluation approach with different exposures, so the subgroup analysis by this approach was equivalent to a meta-analysis of the results of each study. The results reported by Viana et al., were the only ones that overall did not report a negative SMD (SMD = 0.038)  See Tab.Only three papers showed motor function data subsidiary to be analyzed through a correlation coefficient in the meta-analysis ,33,40. Tp = < 0.001, I2 = 75.00%, Tau = 0.181). The global effect was that of a statistically significant negative correlation ; p < 0.001 , were negative. Heterogeneity was very low or null between the different determinations, with the only exception of two determinations between airborne Mn exposure and the results in CATSYS in the section of Tremor HI (dominant and nondominant), presented in the study by Bowler et al. (2016)  See Tab.Seven articles showed motor function data subsidiary to be analyzed through a SMD between groups in the meta-analysis ,39,40,41p < 0.001, I2 = 52.28%, Tau = 0.16). The overall effect was a negative SMD that was statistically significant both under the fixed and random effects model: fixed effect SMD: −0.112; 95%CI (−0.147 to −0.077); p < 0.001, random effect SMD: −0.136; 95%CI (−0.188 to −0.084); p < 0.001 . We obtained similar results regarding the standardized differences of medians: SMD under the fixed effects model: −0.114; 95%CI (−0.149 to −0.079); p < 0.001. SMD under the random effects model: −0.140; 95%CI (−0.194 to −0.086); p < 0.001  = 0.467; p-value (2-tailed) = 0.934.In terms of publication bias in relation to the correlation between cognitive function and exposure to Mn, the funnel plot was asymmetric. When incorporating the ‘Duval and Tweedie (trim and fill)’ procedure , the modp-value (1-tailed) = 0.138; p-value (2-tailed) = 0.276.Regarding publication bias in relation to the cognitive function and exposure to Mn in terms of SMD, the funnel plot was less asymmetric. When incorporating the ‘Duval and Tweedie (trim and fill)’ procedure, no study was included. Therefore, the overall effect adjusted by this procedure was similar to that observed see B. The Egp-value (1-tailed) = 0.058; p-value (2-tailed) = 0.117.As regards to publication bias in relation to the correlation between motor function and exposure to Mn, the funnel plot visually presented a slight asymmetry. However, when incorporating the Duval and Tweedie (trim and fill) procedure, no study was included, so the overall effect adjusted by this procedure was similar to that observed See C. The Egp-value (1-tailed and 2-tailed) < 0.001.Lastly, regarding publication bias in relation to the 71 motor function determinations in terms of SMD, the funnel plot visually presented a slight asymmetry to the left by the base (in those studies with higher standard error). However, when incorporating the Duval and Tweedie (trim and fill) procedure, no study was included, so the overall effect adjusted by this procedure was similar to that observed See D. The EgOur results on cognitive function support a slight but statistically significant negative correlation with environmental airborne exposure to Mn. With concern to the magnitude and interpretation of the correlation coefficient, it should be squared. In this sense, our result (−0.165^2 = 0.0272), indicates that cognitive function and Mn levels share around 2.7% of common variability, and the higher the Mn level, the worse the cognitive function. However, these results are based on only four studies ,41,42,43p < 0.001) in severe (n = 25) (MMSE score < 10) (blood Mn levels 92.08 ± 6.8 μg/L) and moderate cognitively impaired group (n = 86) (MMSE scores range 10–20) (blood Mn levels 77.8 ± 2.4 μg/L) as compared to the age-matched healthy control group (n = 90) (MMSE scores ranged 25–30) (blood Mn levels 52.8 ± 2.8 μg/L). The mildly (n = 72) (MMSE scores range 21–24) (blood Mn levels 64.97± 3.76 μg/L) cognitively impaired group also had significantly elevated levels of Mn compared to the age-matched healthy control group (p < 0.05). As they did not report MMSE scores for exposed and unexposed Mn, these results could not be included in our SMD strategy, but would clearly support a negative association in relation to cognitive function. The results for blood Mn (above the 75th percentile) from the study published by Santos Burgoa et al. (2001) and for air Mn  from the study published by Solís Vivanco et al. (2009), reporting OR > 1 for worse scores on their cognitive tests analyzed, would indirectly support these results [Three articles provided means or medians in the cognitive tests in their respective exposed and non-or-less-exposed populations, so that 26 determinations were subsidiaries of meta-analysis through a SMD ,40,41. T results ,35. Last results .In terms of motor function our results also support the existence of a statistically significant negative correlation of a minor magnitude, based in this case on 36 determinations from five tests reported in three articles ,34,41. TWe found seven articles providing 71 determinations corresponding to eleven different tests, supporting a worse motor function in those with higher environmental exposure to Mn, but also of a very small effect size based on Cohen’s criteria . The stuOne of the main limitations of meta-analysis is the presence of a publication bias . In ordep-value would not be contradictory, in the sense that while a significant p supports the existence of bias, a nonsignificant p would not rule it out as it can be explained for example by a lack of statistical power. Nevertheless, by including studies from the left, the publication bias, if any, would favor the hypothesis of a larger negative correlation, although the procedure itself in this case shows that the change in the overall adjusted effect is not relevant. Regarding the SMD between groups for cognitive function, the funnel plot visually did not present asymmetry and no study was included when incorporating the Duval and Tweedie (trim and fill) procedure. All this supports the validity of our results, in the sense that there is no publication bias, or at least not in the way that the impact on cognitive function is considerable.Regarding the publication bias in the correlation between cognitive function and exposure to Mn, the funnel plot visually presented asymmetry, including studies on the left when incorporating the Duval and Tweedie (trim and fill) procedure, which however had little impact on the overall adjusted effect . The Eggp value of 0.058 in the unilateral contrast test. However, the Duval and Tweedie (trim and fill) procedure, did not incorporate any studies. For the SMD, the asymmetry in the funnel plot was associated with an Egger T test with very significant p values (p < 0.001), which would support the existence of a bias, but which, however, did not correspond either with the inclusion of any study using the Duval and Tweedie (trim and fill) procedure. The same occurred in the sensitivity analysis by excluding the “Sniffin’ sticks test” determinations. Thus, there would also be no conclusive evidence of the existence of a publication bias in motor function, also supporting the validity of our estimates.In relation to the motor function and correlation approach, the funnel plot visually presented a slight asymmetry, and the Egger test marked the intercept value at −1.50; with a In addition to the possibility of publication bias, another limitation of the present meta-analysis would be the low quality of some primary studies, and the small sample size of most of them. Regarding the design and quality of the primary studies, all studies included in our systematic review had a cross-sectional design, with cognitive or motor functions assessed at the same time as exposure, i.e., in none of the designs there was a follow-up period to assess the impact of short, medium or long-term exposure on motor or cognitive functions. This reinforces the need for prospective studies in adults, with a follow-up period in which the appropriate time sequence between exposure (cause) and effect on cognitive and motor functions should be preserved. In relation to the control of bias such as confounding, two studies attempted to control for this bias in the design phase by matching exposed and unexposed groups for gender and as closely as possible, for age and level of schooling ; and cogn = 183) and healthy controls (n = 90) are particularly high in comparison with those of other studies , which could be explained by a selection bias  or by a bias in the characterization of the exposure . Overall, this study obtained the lowest quality among analyzed in our quality approach, and it was the only one obtained a grade of “poor”. This study contributed only with a single determination in the meta-analysis section of correlation and cognitive function. By removing this determination in a sensitivity analysis, the results did not change; so it was decided to include it, presenting the overall results of the four studies providing 56 determinations in the meta-analysis.Regarding the study published by Iqbal et al., in addition to the lack of use of multivariate analysis, it should be noted that the reported blood Mn levels in cognitive impairment volunteers (2 > 50%), since under the random effects model the studies providing greater heterogeneity would have a lower relative weight. Nevertheless, due to the a priori sources of heterogeneity identified, it was considered appropriate to additionally show the results under both models, at least in the presented figures, even for those strategies with low heterogeneity.In relation to the choice of model for meta-analysis, the heterogeneity of the results of the studies in the different analyses was variable, suggesting the choice of a fixed or random effects model for the synthesis of results adapted to the specific heterogeneity for each particular analysis strategy, showing the results under the random effects model in the case of moderate & high heterogeneity  [3 [3 in PM10 and 50 ng/m3 in PM2.5, have been reported in Marietta [3 in PM10 and 30 ng/m3 in PM2.5 in East Liverpool, Ohio, US [3 have been measured using personal exposure PM10 samplers [3 in PM2.5 was measured in Simões Filho (Brazil), the same area considered by Viana et al. (2014) study [Several studies included in this systematic review have been conducted in areas near air Mn sources, such as Mn ore mines , Mn ore 3 , the US l basis) , or the asis) [3 . Thus, mMarietta ,36,37,42Ohio, US ,42. In nsamplers ; and a m4) study , in an e4) study .3 [At the moment, since no human biomonitoring values derived on the basis of toxicological and epidemiological studies are available for Mn in human matrices commonly used as biomarkers of exposure, such as whole blood, hair, nails or saliva , there i3 ,56. The In conclusion, our meta-analysis shows a statistically significant negative correlation between cognitive and motor functions . Regarding the SMD approach, our results also support worse cognitive and motor functions in the exposed, although only for motor function statistical significance was obtained. This would support the need to regulate airborne Mn levels. However, further follow-up studies with a mostly systematic methodology, using a standardized, more homogeneous battery of neurological/neuropsychological tests are needed to deepen the effects derived from environmental airborne Mn exposure and their clinical relevance.


Postoperative serum albumin  was a significant predictor for AKI. Age  and ACE inhibitors  were associated with increased the risk of AKI, but the results were not statistically significant. No significant publication bias was identified through statistical tests . In conclusion, our findings indicated that the pooled AKI following hip fracture surgery was approximately 17%. Postoperative serum albumin was a potential significant risk factor for AKI.Acute kidney injury (AKI) was a frequent complication following hip fracture surgery, but recent studies reported inconsistent findings. Our study was aimed at clarifying the prevalence and risk factors of AKI after hip fracture surgery. Pubmed, Embase, and Web of Science were systematically searched from the inception to March 2020 to identify observational studies investigating the prevalence and risk factors of AKI in patients undergoing hip fracture surgery. Pooled prevalence and odds ratios (ORs) with 95% confidence intervals (CIs) were estimated using a random-effects model. Publication bias was evaluated with a funnel plot and statistical test. All the statistical analyses were performed using Stata version 12.0. A total of 11 studies with 16,421 patients was included in the current meta-analysis. The pooled prevalence of AKI in patients undergoing hip fracture surgery was 17%  with substantial heterogeneity (I  Acute k, ,5. Notab, ,7. ThereActually, the prevalence of AKI after hip fracture surgery was inconsistent across several published studies, ranging from 5% to 60%. This controversiality may originate from the heterogeneous populations, different AKI definition, and inconsistent follow-up durations. For clinicians, an inaccurate estimation of AKI prevalence will help them comprehensively understand an overview of the disease burden and evaluate whether specific treatment strategies are effective for preventing AKI. Additionally, increasing studies also revealed that some risk factors were significant predictors for AKI after hip fracture surgery, including chronic kidney disease, postoperative serum albumin, and intraoperative hypotension . NeverthConsidering the aforementioned controversy and uncertainty, we performed a meta-analysis and systematic review to systematically investigate the prevalence and risk factors for AKI in patients undergoing hip fracture surgery, which contribute to providing the available optimum epidemiological evidence on this topic.This meta-analysis was undertaken according to the guideline of the Meta-analysis of Observational Studies in Epidemiology (MOOSE) checklist and the Preferred Reporting Items for Systematic Reviews and Meta-Analysis (PRISMA) statement ,10.Pubmed, Embase, and Web of Science were systematically searched from inception to March 2020 to identify observational studies that reported the prevalence and risk factors of AKI in patients undergoing hip fracture surgery. The search strategy was established using the terms of ‘acute kidney injury’, ‘hip fracture surgery’ and their variants. Also, the references of included studies and some important reviews were manually checked for any potential inclusion.Studies involving the following inclusion criteria were included in the current meta-analysis:(1) observational studies, including cohort studies, case-control studies, or cross-sectional studies; (2) studies investigating the prevalence and risk factors of AKI in patients undergoing hip fracture surgery. Only studies published in English were considered. The Retrieved studies were individually evaluated for eligibility by the two investigators with discrepancies resolved through discussion.We applied the pre-designed table to extract the following information: first author, publication year, study period, country, operation type, case number, the number of patients with AKI, AKI definition and risk factors of AKI. The primary outcome was the prevalence of AKI after hip fracture surgery. The secondary outcomes were risk factors of AKI after hip fracture surgery. Furthermore, only ORs with 95%CIs on multivariate analysis in the included studies were extracted. The quality of the included studies was evaluated using Newcastle–Ottawa Scale (NOS) score. The sco2 statistic was applied to evaluate the statistical heterogeneity across eligible studies and I2 > 50% was regarded as substantial heterogeneity [The prevalences of AKI after hip fracture surgery were extracted from included studies. Pooled prevalences with 95% confidence intervals (CIs) were calculated using the generic inverse-variance method. The correlative dimension of risk factors with AKI was estimated as odds ratios (ORs) with 95%CIs using a random effect model when considering substantial statistical heterogeneity across the included studies. Only candidate risk factors reported in two or more eligible studies on multivariable models were considered for meta-analysis. Iogeneity ,13. Metaogeneity ,15. A twThe current meta-analysis was aimed at exploring the prevalence and risk factors of AKI after hip fracture surgery. Pubmed, Embase, and Web of Science were systematically searched to identify observational studies investigating the prevalence and risk factors of AKI in patients undergoing hip fracture surgery. We found that a total of 11 studies with 16,421 patients were included in the current meta-analysis. The pooled prevalence of AKI in patients undergoing hip fracture surgery was approximately 17%. Postoperative serum albumin was a significant predictor for AKI. Age and ACE inhibitors were also associated with increased the risk of AKI, but the results were n’t statistically significant.A total of 810 items were identified through systematically searching three databases. Also, another 24 items were obtained from other sources. After removing duplicated and irrelative items, the full texts of 57 remaining articles were screened for possible eligibility. Eventually, a total of 11 studies with 16,421 patients was included in the current meta-analysis . The flo2 = 95%) . The statistical heterogeneity was significant (I2 = 95%) . Subsequ2 = 95%) . We furtWe also explore the potential risk factors associated with AKI after hip fracture surgery. A total of seven studies reported AKI-associated risk factors on multivariate or adjusted model  3,6,8,1,19,21,22The current meta-analysis revealed that AKI was a relatively frequent complication in patients undergoing hip fracture surgery with pooled prevalence ranging from 14% to 21%. Additionally, postoperative serum albumin was identified to be a significant risk factor for AKI following hip fracture surgery.The current meta-analysis based on 11 observational studies indicated that the overall pooled prevalence of AKI following hip fracture surgery was 17% with substantial heterogeneity. Considering that the significant heterogeneity may impair the credibility of the pooled estimate, meta-regression was performed to explore the potential sources of statistical heterogeneity. Furthermore, we found that publication time, sample size, and NOS score may not be responsible for significant statistical heterogeneity. Subsequently, we conducted subgroup analysis and sensitivity analysis to explore the prevalence of AKI in sub-population. Interestingly, the results of subgroup analysis and sensitivity analysis were basically consistent with the overall pooled effect, which suggested that the overall pooled estimate was robust and reliable. A previous meta-analysis showed that the overall estimated prevalence rates of AKI in patients undergoing total hip arthroplasties are 6.3%. ObviousThe current study is the first meta-analysis which systematically investigated the prevalence and risk factors after hip fracture surgery. An important strength of our study lies in its accordance MOOSE checklist and PRISMA guidelines. Two reviewers independently conducted literature search, data extraction, quality assessment, and statistical analysis, which facilitate the transparency and replicability of the meta-analysis. There also existed several limitations in the current study. First, our meta-analysis showed substantial statistical heterogeneity, which may potentially impair the reliability of the pooled estimate. Subsequently, we performed meta-regression to explore the sources of statistical heterogeneity and none of the significant factors was identified to be responsible for significant heterogeneity. A possible interpretation is that multiple clinical and methodological differences across included studies, but not individual factors, contribute to the significant statistical heterogeneity. Irrespective of the statistical heterogeneity, the results of subgroup analysis and sensitivity analysis were basically consistent with the overall pooled effect, which showed the robustness and reliability of the pooled estimate. Second, we evaluated the potential publication bias using the funnel plot and statistical tests. The statistical results showed that the publication bias was not statistically significant, but the funnel plot seemed to be asymmetric. Considering the inconsistency, the potential publication bias still cannot be excluded, although we performed a systematic literature search in the meta-analysis. Third, some risk factors reported in included studies were not pooled for meta-analyses owing to limited studies, which may bias the authentic effects for AKI. The pooled analysis based on two studies found that ACE inhibitors may not be a significant risk factor for AKI following hip fracture surgery. Actually, many studies found that ACE inhibitors were a significant predictor for AKI . Also, mThe current meta-analysis revealed that the pooled AKI in patients undergoing hip fracture surgery was approximately 17%. Postoperative serum albumin was identified as a potentially significant risk factor for AKI. Further high-quality studies should be warranted to systematically clarify the prevalence and risk factors of AKI following hip fracture surgery.


Subsequent association analysis revealed significant negative correlations between plasma free radical scavenging nutrients and plasma ox-LDL levels, indicating a general benefit of PMS in alleviating oxidative stress by providing exogenous oxidant scavengers.A meta-analysis has been widely applied to draw general conclusions using a set of studies with similar purposes and designs. This study aimed to perform a meta-analysis of six randomized placebo-controlled trials, independently conducted for the relationship between a plant-based multivitamin/mineral supplementation (PMS) and oxidative stress for 6 to 8 weeks, to provide overall estimates of those effects. In detail, linear mixed model analysis was first conducted on each study to obtain individual estimates; then, two types of meta-analysis were applied to combine the individual estimates from all available studies  and region-specific studies . In the meta-analysis, we selected 19 biomarker variables that overlapped in at least two studies and found 6 variables significant in at least one meta-analysis. The overall estimates of beta coefficients were 0.17 for vitamin C, 0.80 for vitamin B Oxidative stress is an imbalance between the generation of reactive oxygen species (ROS) and the ability to deactivate them. The association of oxidative stress with many chronic diseases has been extensively studied and found to be an essential factor in inducing public health problems in developed and developing countries . TherefoSuch consistency and inconsistency have also been found in randomized and placebo-controlled trials (RCTs), which were independently conducted to evaluate the effects of the same plant-based multivitamin supplement (PMS) on reducing oxidative damage, maintaining endogenous ROS homeostasis, or providing heart health benefits in general. The PMS contained nutritional doses of multiple vitamins/minerals and phytonutrients from extracts or powders of various fruits and vegetables. Similar protocols were used but on regionally different populations. The directions of the results were generally consistent; however, not every individual study detected statistical significance, probably due to the relatively small sample size . In addip-value. Thus, it is expected to provide better estimation results than single studies. However, conducting a meta-analysis is impossible if the extracted studies do not contain enough information of parameter estimates, such as beta coefficients or standard error of interest [Meta-analysis is a set of statistical procedures synthesizing data by combining results from independent studies to integrate the findings and provide overall estimates of the relationship between variables . It has interest .Therefore, the current study aimed to perform a meta-analysis based on the previously performed RCTs of PMS to shed light on the controversy and reach a general conclusion about its effectiveness in maintaining health by reducing oxidative stress. In detail, we collected individual subject datasets of six  and 10 minerals . In addition, phytonutrients were provided from the fruits and vegetables, including acerola, alfalfa, apple, bilberry, black currant, blueberry, brassica, broccoli, carrot, citrus, cranberry, elderberry, grape, grapefruit, horseradish, kelp, lemon, mandarin orange, marigold, onion, orange, parsley, peppermint, plum, rosemary, spinach, tomato, turmeric, and watercress. The placebo samples were formulated to match the shape and color of the PMS sample, primarily using dextrose, microcrystalline cellulose, silicon dioxide, croscarmellose sodium, and magnesium.The PMS sample provided nutritional doses of 14 vitamins  or the International Clinical Trials Registry platform of the WHO before recruitment began. Principal investigators of individual RCTs willingly provided the requested raw data and related information for independent statistical reanalysis.Six independent RCTs investigating the effectiveness of PMS against oxidative stress were collected for this meta-analysis . This meDespite similar aims and design, the six RCTs vary to a more or less considerable extent in the variables used in each study. Therefore, we evaluated data from the six studies and selected variables for meta-analysis based on the degree of overlap among studies. We only used the variables overlapped in two or more studies for meta-analysis and the corresponding interpretation of the results .Data preprocessing was conducted before the meta-analysis. First, we removed data that were not complete, such as “Quantity Not Sufficient (QNS).” Second, inequalities to mark specific cutoffs (ex. “> 500”) were substituted differently depending on the existence of observations beyond the cutoff for the particular variable. If there were no observations beyond the value in the inequality (ex. “>500” is the maximum observation) for a variable, that value was imputed in place of the inequality. When there were observations beyond the cutoff presented by the inequality, the imputed value was the mean of values within the inequality criteria before the following cutoff criterion. For instance, if there were two inequalities, “>500”, and “>600”, as well as observations 501, 503, 532, 640, and 700, “>500” would be imputed as 512 , while “>600” would be imputed as 670 (mean of 640 and 700). Finally, to analyze the intervention effects of PMS during the experimental period, we removed data without measurements at the start or end of the trial duration (“unpaired data”) and included only the paired data in the analysis.k is the rank of the dataset and n is the sample size of the dataset from the relevant individual trial. The “inverse normal” transformation was used to reduce the skewness and effects of outliers, which could not be achieved through square-root transformation. After the transformation step was finished, we also performed a standardization step on the transformed expression levels to make effects of PMS from the individual studies comparable and, consequently, to combine the effects in the meta-analysis.Then, we additionally conducted a normalization process, which included a data transformation step, to reduce skewness and improve normality , and a s(n+0.25) . Then, wi  denotes subject id, j  denotes the time point of measurements, and k  denotes the index of individual studies. Thus, for each variable, ith subject, at the jth time point, and in the kth study. Among the explanatory variables, wk is a binary week status, The univariate analyses were first conducted to assess PMS effects based on different variables. To this end, the following linear effect model was used to handle repeated measures, as presented in Equation (1):p-value. Then, we conducted Cochran’s Q test [ith individual trial, represented as the reciprocal of the standard deviation obtained from the individual trials, ith individual trial, k is the number of trials. If the test p-value is <0.05, the degree of heterogeneity is regarded as significant. In addition, the I2 statistic [The values of s Q test  to choostatistic  was usedFinally, we conducted an overall and subgroup meta-analysis. Among the six studies included in this meta-analysis, four were performed in Asian subgroup meta-analysis, and the other two were in Western subgroup analysis, such as the USA and Russia. We categorized the studies as above because, as Asians accounted for a low percentage of the subjects studied in the USA (5.6%) and Russia (6.5%), it seemed reasonable to group the USA and Russia into the Western subgroup. In addition, because Korean studies account for two of the four Asian studies, a subgroup was created. Therefore, the subgroups are listed as follows: (1) Korean (2013 Korea and 2019 Korea), (2) Asian , and (3) Western (2015 USA and 2018 Russia).p-value ≤ 0.05 was considered statistically significant. The “lmer” function in the R package “lmerTest” was used to fit the linear mixed model. The heterogeneity test and meta-analysis were conducted using the “rma” function in the R package “metafor,” which can analyze random-effect (using restricted maximum likelihood estimator) and fixed-effect (using inverse-variance weighting method) analyses. When the interaction term between treatment and week of a variable was statistically significant in the overall group or the subgroup meta-analysis, it was regarded to have a significant effect induced by PMS and we interpreted its biological meaning.All statistical analyses were performed using R 4.0.2, and a n = 305) of the sample was male.The characteristics of trials and subjects of the six studies included in this meta-analysis are summarized in β-carotene was the most shared variable by being included in all studies, followed by homocysteine included in five studies; vitamin C, B6, B12, folate, and C-reactive protein (CRP) included in four studies; lutein, oxidized-low-density lipoprotein (ox-LDL), and Comet assay included in three studies; and α-tocopherol, γ-tocopherol, zeaxanthin, α-carotene, lycopene, malondialdehyde, Comet assay with H2O2 challenge, 8-hydroxy-2′-deoxyguanosine, and Short Form 36 questionnaire included in two studies. The selected variables were classified into three categories: oxidative stress, chemical scavenger, and quality of life.12 and β-carotene were normalized by inverse-normal transformation, while those for vitamin C, vitamin B6, folate, and ox-LDL were normalized by square-root transformation, based on the rule for the normalization process explained in the Materials and Methods section.6 , vitamin B12 , folate , and β-carotene . Therefore, we employed a random-effect model for the meta-analysis of these variables. In contrast, a fixed-effect model was applied for the meta-analysis of the other variables, considering the homogeneous effects of PMS on them. Additional statistical heterogeneity evaluation was performed using the I2 statistic, which measures the relative magnitude of heterogeneity [2 > 50%, confirming substantial heterogeneity. Although one exception was observed in the overall meta-analysis of vitamin C with Cochran’s Q p-value = 0.095 and I2 = 52%, the other variables employing the fixed-effect model showed I2 < 50%. However, it was notable that the effects of vitamin C were homogeneous in the overall group and subgroups, based on the Cochran’s Q test.Substantial heterogeneity existed in the meta-analyses of vitamin Bogeneity ,22. As ep = 0.008). Studies 2 and 4 also showed positive effects but did not reach statistical significance. Study 1 reported a contradictory effect, although its effect size was very small, without statistical significance (p = 0.6842). Therefore, the overall meta-analysis result remained marginally significant (p = 0.0797). Similar trends were found for the impact of PMS on changing plasma vitamin B6 and B12 levels. For vitamin B6, Studies 2, 4, and 5 all showed significantly positive results , while Study 1 showed a non-significant negative result. Consequently, the overall meta-analysis result was found to be positively significant (p = 0.0097). For vitamin B12, Studies 2 and 4 showed significantly positive results , while Study 5 showed a tendency of positive effect and Study 1 a non-significant negative result. As a result, the overall meta-analysis result was found to be positively significant (p = 0.0347). In the case of folate, the most consistent positive results were observed, with Studies 2 (p < 0.0001), 3 (p < 0.0001), 4 (p = 0.0869), and 5 (p = 0.007) being accounted for. Thus, the overall meta-analysis resulted in very large effect size (p = 0.0001). β-carotene is the most shared variable, but the results were inconclusive. Study 1 showed a non-significantly negative change, Studies 3 and 5 non-significantly positive changes, Study 6 a tendency of increase, and Studies 2 and 4 significant increase . However, when the data were all combined for meta-analysis, the resulting beta coefficient and p-value indicated a globally significant positive effect of PMS on plasma β-carotene level (p = 0.0307). Finally, PMS influence on plasma ox-LDL level was non-significantly negative in Studies 3 and 6, while non-significantly positive in Study 5. However, surprisingly, the p-value was improved when data were combined for the overall meta-analysis.6 , B12 , folate , and β-carotene  with plasma ox-LDL , vitamin B6 (p = 0.0006), vitamin B12 (p < 0.0001), and folate (p = 0.0024) were found to be positive and significant in the Western subgroup, the Asian subgroup showed a significant positive influence only on the plasma folate level (p < 0.0001). In the meantime, the Korean subgroup showed significant effects of PMS on β-carotene (p = 0.0263) and Ox-LDL (p = 0.0489).The overall meta-analysis results support the positive influence of the PMS to increase vitamin C, B6, B12, folate, β-carotene, and ox-LDL, while the univariate analysis in individual studies found conflicting results. (2) The antioxidant potential of PMS may be derived from the increase in the plasma levels of free radical scavenging vitamins, including vitamin B6, B12, folate, and β-carotene. (3) There were some regional differences in the sensitivity to PMS supplementation. Likewise, our results supported the general knowledge of how meta-analysis could overcome the lack of power, settle controversies arising from studies with conflicting findings, and derive conclusions not directly addressed by individual research [This meta-analysis of six RCTs comprising 817 PP participants from different regional populations helped forge a better understanding of the effectiveness of PMS in improving oxidative defense compared to the individual RCTs. The main findings are the following: (1) Our meta-analysis could conclude that PMS remarkably influenced plasma vitamin C, Bresearch ,24.When we ran a scan on “meta-analysis for multivitamins” on the PubMed search engine, we found that there were 36 published papers. The most frequent issue was about the benefits of multivitamin use during pregnancy and prenatal development (18 cases), followed by those related to risk reduction in chronic diseases, such as cancers (4 cases), circulation diseases (4 cases), eye diseases (4 cases), cognitive diseases (2 cases), infections (2 cases), hip fracture (1 case), and mortality (1 case). Unfortunately, these previous meta-analyses estimated a clinical endpoint or surrogate marker that is more difficult to observe than clinical markers, limiting the success of meta-analysis . Unlike 6, B12, folate, and β-carotene, and in suppressing oxidative stress, as indicated by reduced ox-LDL in plasma. Subsequent association analysis suggested that increased plasma levels of antioxidant nutrients were involved in suppressing oxidative stress, based on a confirmed correlation between plasma vitamins and ox-LDL levels. However, vitamin C was an exception in this study. This result is consistent with a previous study that reviewed the relationship between dietary vitamin C intake and plasma levels of vitamin C [Overall, our meta-analysis results showed that PMS might have a beneficial effect in enhancing free radical scavenging nutrient levels in plasma, as shown in the cases of vitamin C, Bitamin C . It was itamin C .2 statistic. Here, statistical heterogeneity refers to variation in the true effect sizes between the studies, rather than the variation due to sampling error [6, B12, folate, and β-carotene, and thus we applied random-effect models to calculate pooled estimates of the outcomes. Although this approach usually makes little difference to the final estimate, the overall meta-analysis resulted in large effect sizes for vitamin B6, B12, folate, and β-carotene. In the meantime, vitamin C and ox-LDL had a relatively low heterogeneity in the overall meta-analysis. We thus applied fixed-effect models, in which the pooled effect estimates the common true effect size.The strength of our study is worth mentioning. Most importantly, unlike meta-analysis based on aggregating study data, our meta-analysis has the advantage of using individual subject data from independently performed RCTs, which are often considered the gold standard for meta-analysis . Individng error . SubstanOn the other hand, our study also has some limitations. Multivitamin supplements from natural sources tend to be preferred over many other nutritional supplements distributed in the market, probably due to the presence of cofactors known as phytonutrients . HoweverIn conclusion, our meta-analysis supports a general benefit of PMS in alleviating oxidative stress by providing free radical scavenging nutrients in the general population with a habitually low intake of fruit and vegetables. In future studies, we will adopt metabolomics technologies to include the analysis of phytochemicals and their metabolites in plasma and expand the concept of the benefits of plant-based vitamin supplementation.


CTLA-4) gene polymorphisms may be involved in the risk of Rheumatoid arthritis (RA). However, evidence for the association remains controversial. Therefore, we performed a meta-analysis to confirm the relationship between CTLA-4 gene polymorphisms and RA. The pooled odds ratios (ORs) and 95% confidence intervals (CIs) were calculated to assess the strength of association. Stratified analysis was conducted by ethnicity. In total, 66 case-control studies including 21681 cases and 23457 controls were obtained. For rs3087243 polymorphism, significant association was detected in Asians  and Caucasians . For rs231775 polymorphism, significant association was observed in the overall , and in Asians , but not in Caucasians. However, there was no association between rs5742909 polymorphism and RA. This meta-analysis confirmed that rs3087243 and rs231775 polymorphism were associated with the risk of RA in both overall population and ethnic-specific analysis, but there was no association between rs5742909 polymorphism and RA risk.Cytotoxic T lymphocyte-associated protein 4 ( Rheumatoid arthritis, one of the most common inflammatory joint diseases in humans, is characterized by inflammation in synovium, destruction of cartilage and bone, generation of autoantibody, and complications of systemic organs . AlthougCTLA-4), one of widely studied non-HLA susceptibility gene of RA, is mainly expressed on the surface of Treg cells and conventional T cells and suppresses self-reactive T cell responses via downregulating ligand availability for the costimulatory receptor CD28 to elicit inhibitory signals [CTLA-4 have already been proved to be candidates of the risk of the common autoimmune diseases at the genetic level [CTLA-4 plays a vital role in regulating T cell function [CTLA-4 expression or function is most likely associated with the pathogenesis of RA. Single nucleotide polymorphisms in the CTLA-4 gene may contribute to abnormal levels of CTLA-4, and subsequently play a leading part in the susceptibility to RA [Cytotoxic T lymphocyte-associated protein 4 , -318C/T (rs5742909) and CT60 G/A(rs3087243), are most-often studied for the association with the predisposition of RA [CTLA-4 genetic polymorphisms and the risk of RA has been assessed in several meta-analyses [CTLA-4 with susceptibility to RA. Here we use the latest case-control data to carry out an updated and comprehensive meta-analysis and obtain a more accurate estimation of the effect of the 3 SNPs (+49A/G (rs231775), CT60 G/A(rs3087243) and -318 C/T (rs5742909)) on RA risk.Among the identified SNPs in this gene, these three loci of on of RA . Howeveranalyses –23, someanalyses , 24–27. Based on the predetermined inclusion criteria, 66 eligible case–control studies with 42 articles were enrolled ultimately in the current analysis , 24–56. CTLA-4 CT60(rs3087243) SNP was observed in all genetic comparisons  (P=0.001) and the homozygote comparison . The heterozygote model and dominant model detected also this correlation in Latin Americans and the allelic comparison detected this correlation in Africans, but both the two populations needed more enrolled studies to elevate statistical power because this analysis currently included individually only one study. The outcomes were shown in I2 > 50% and P>0.1, the Fix-effect model was used for the synthesis; otherwise, the Random-effect model was used.By analyzing quantitatively allele or genotype distribution of 16394 patients and 17453 controls, a significant association between RA and =0.0003) . Among tCTLA-4 +49A/G (rs231775) SNP. The overall pooled ORs of all the populations were as follows: G vs. A: OR =1.16, 95% CI =1.08-1.25, P<0.0001; GG vs. AA: OR =1.29, 95% CI =1.12-1.50, P=0.0006; GA vs. AA: OR =1.19, 95% CI =1.07-1.32, P=0.001; GG + GA vs. AA: OR =1.24, 95% CI=1.11-1.39, P=0.0001 and GG vs. GA+AA: OR =1.15, 95% CI=1.02-1.30, P=0.02. The main results of overall analyses were shown in G vs. A: OR =1.27, 95% CI =1.10-1.47, P=0.001; GG vs. AA: OR =1.58, 95% CI =1.24-2.01, P=0.0002; GG + GA vs. AA: OR =1.33, 95% CI=1.17-1.51, P<0.0001; GG vs. GA+AA: OR = 1.15, 95% CI =1.02-1.30, P=0.02). In Latin American population, rs231775 SNP was a significant risk factor of RA, but it only included single study and the result might be incredible. Besides, no association of the rs231775 SNP with RA risk was found among the Caucasian population in all genetic comparisons when the Elshazli’s study [G vs. A: OR =1.07, 95%CI =0.99-1.15, P =0.08; GG vs. AA: OR = 1.07, 95% CI = 0.92–1.23, P=0.37; GA vs. AA: OR = 1.15, 95% CI =1.00-1.31, P=0.05; GG + GA vs. AA: OR =1.14, 95% CI=1.00-1.29, P=0.05 and GG vs. GA+AA: OR =1.00, 95% CI=0.90–1.11, P=0.98)  . There wT vs. C: OR =1.21, 95% CI =0.93-1.57, P=0.15; TT vs. CC: OR =1.71, 95% CI =1.08-2.73, P=0.08; TC vs. CC: OR =1.19, 95% CI =0.84-1.69, P=0.33; TT+TC vs. CC: OR =1.19, 95% CI=0.84-1.69, P=0.33 and TT vs. TC+CC: OR =1.43, 95% CI=0.90-2.27, P=0.13)  . Meanwhi P=0.13) . As the 2I) and found that heterogeneity existed in some genetic models of rs231775 SNP and rs5742909 SNP (2I>50%). On the other hand, sensitivity analysis was adopted to evaluate the effect of a single study on the pooled ORs by removing each study in turn from the pooled analysis. Although the heterogeneity had not changed obviously, the P values for pooled ORs under allelic comparison, heterozygous comparison and dominant comparison were reversed when the study [To ensure the reliability of the results, we first evaluate the heterogeneity . These CTLA-4 polymorphism on RA risk [CTLA-4 with RA was contrary to the others; the genetic models which indicated significant association were diverse in these analyses. These differences were mainly originated from divergent diagnostic criteria, limited number of studies and sample sizes. On the other hand, all these meta-analyses focused on only one of the three well-studied loci except Li’s study [CTLA-4 gene and RA susceptibility.It should be pointed out that previous several meta-analyses have summarized the effect of  RA risk –23, 65. ’s study , 24–27 hCTLA-4 gene had various SNPs, the function of protein CTLA-4 depended on the whole gene and RA was a multigene susceptibility disease, more SNPs of CTLA-4 should be included. Thirdly, certain degree of heterogeneity still existed in rs5742909 polymorphism and some genetic models. Although the elimination of each single study did not distinctly alter the P value, the results must still be treated cautiously. Fourthly, inadequate raw data in some studies result in the inability to calculate the number of the genotypes and perform stratified analysis by age, gender and autoantibody status such as RF etc. As a consequence, any potential gene-environment and gene-gene interactions could not be accessed.However, some limitations of our study should be acknowledged. Firstly, the small sample size in some studies and the limited studies for some stratified analysis were not sufficient enough to detect the relationship. Especially, the results of populations including only one study should be interpreted with caution. Secondly, we only investigated the role of three loci polymorphisms. As CTLA-4 played in the pathogenesis of RA.In conclusion, this meta-analysis suggested that rs3087243 polymorphisms were corelated with a reduced RA risk in both Asian and Caucasian populations, rs231775 polymorphisms was associated with an increased risk of RA in Asians, and rs5742909 polymorphism had no significant association with RA risk. Larger-scale studies of populations with different ethnicities are encouraged to validate the role This meta-analysis was performed according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines .CTLA-4 or cytotoxic T lymphocyte antigen-4” and “single nucleotide polymorphism or polymorphism or variant or variation” and “rheumatoid arthritis or RA”. The bibliographic lists of included studies were also browsed for potential related studies. There were no restrictions on language and publication date in this study.From the databases PubMed, EMBASE, Web of Science and, the China National Knowledge Infrastructure (CNKI) and Wan Fang data, a comprehensive systematic literature retrieval was conducted to derive all relevant studies published before 10 October, 2020 (the search was constantly updated to submission). The following terms as Medical Subject Heading and free words were applied: “CTLA-4 polymorphism and RA risk; 3) with sufficient data for extract odds ratios (ORs) and 95% confidence intervals(CIs); (4) with reported allele or genotype numbers or frequencies in cases and control group; 5) with a clear diagnostic criteria. Accordingly, we excluded meaninglessness literatures if they had the following trait: 1) case report, comment, animal studies and conference abstracts; 2) with no detailed allele or genotype data; 3) duplications or no controls.The current meta-analysis used the following inclusion criteria to screen available literatures: 1) case-control study; 2) evaluation of the associations between Two independent investigators respectively conducted a literature search according to the above search strategy, screened each article based on the predesigned inclusion and exclusion criteria, and extracted data from these eligible studies. It would be settled by discussion with the third party when the disagreement between investigators occurred. The following information was collected from every paper: 1) first author's surname, 2) the year of publication, 3) country or region of origin, 4) ethnicity, 5) total numbers of cases and controls, 6) genotype method, 7) diagnostic criteria, 8) polymorphism locus, 9) allele distribution or/and genotype distribution.http://www.ohri.ca/programs/clinical_epidemiology/oxford.asp). Any divergence between two investigators was solved by discussion until agreement was reached.The methodological quality of included studies was accessed in light of the Newcastle–Ottawa Scale (NOS) for the evaluation of observational studies . In brieI2 values which range from 0% to 100%. 25%, 50%, and 75% were regarded as respectively low, moderate, and high level [I2 was more than 50%. If not, the fixed effect model was employed. Hardy–Weinberg equilibrium (HWE) was tested in the control group for all studies by Chi-square test to judge whether the selection bias existed. Potential publication bias was examined by funnel plots. Besides, the current meta-analysis had carried out subgroup analyses by the racial descent to assess the effects of ethnic background.The strength of association of rs231775, rs5742909 and rs3087243 SNPs with RA risk was appraised via estimating ORs with their corresponding 95% CIs. For each SNP, the pooled ORs were calculated individually for five gene models . The Z test was used to evaluate the significance of the pooled ORs. p<0.05 was judged as statistically significant difference. Statistical Heterogeneity between studies was assessed by Chi square and gh level , 69. TheThe above statistical analyses were performed using Review Manager 5.3 software . All the P values were 2-sided and P<0.05 signified statistically significance.Supplementary Table 1


Retinol binding protein 4 (RBP4) has been proposed to play a role in the pathophysiology of coronary artery disease (CAD), but previous findings on the association of RBP4 levels with CAD are inconsistent.I2 > 30%) and data with low heterogeneity were analysed using a fixed-effects model (I2 ≤ 30%). Moreover, a bias-adjusted quality-effects model was generated, and the prediction interval was also calculated under the random-effects model.A meta-analysis based on observational studies was conducted to evaluate the association between circulating RBP4 levels and CAD. Databases including PubMed, Web of Science, Embase, Google Scholar and ClinicalTrials.gov database were searched for eligible studies published up to 12 July 2021. Standard mean differences (SMDs) with 95% confidence intervals (CIs) were calculated using the inverse variance heterogeneity (IVhet) and random-effects model for data with moderate and high heterogeneity (I2: 96.00%). The quality-effects model produced consistent results. However, the association turned to be significant under the random-effect model , whereas the 95% predictive interval (PI) included null values (95% PI: − 0.82-1.74). Subgroup analyses illustrated a positive relationship between CAD and RBP4 levels in patients with complications . The meta-regression analysis revealed that the mean BMI of patients (P = 0.03) and complication status (P = 0.01) influenced the variation in SMD.Two nested case-control studies, one cohort study and twelve case–control studies with a total of 7111 participants were included. Circulating RBP4 levels in patients with CAD were comparable to those in the controls under the IVhet model (SMD: 0.25, 95% CI: − 0.29-0.79, There was low-quality evidence that patients with CAD exhibited similar circulating RBP4 levels compared with controls, and high inter-study heterogeneity was also observed. Thus, RBP4 might not be a potential risk factor for CAD. Comparisons among different subtypes of RBP4 with larger sample size are needed in the future.The online version contains supplementary material available at 10.1186/s12944-021-01516-7. Coronary artery disease (CAD), the most frequent cardiovascular disease, is the leading cause of mortality and morbidity worldwide . In 2017RBP4, a plasma transport protein that delivers retinol from liver to the tissues, may play an important role in CAD through its involvement in the progression of inflammatory mechanisms in adipose and vascular tissues . Moreovehttps://www.crd.york.ac.uk/prospero/) with number CRD42020152286. A systematic literature search was performed in four different databases including PubMed, Embase, Web of Science, Google Scholar and ClinicalTrials.gov database for published and unpublished literature using the following keywords with respect to specific search tips of each database:  AND . The details of search strategy were shown in Supplementary Table The present meta-analysis was conducted in accordance with the guidelines of Meta-analysis of Observational Studies in Epidemiology  and PrefThe selection criteria were listed below: (1) case–control, nested case–control or prospective cohort studies that investigated the relationship of the circulating RBP4 concentration with the risk of CAD; (2) mean circulating RBP4 level and its standard deviation, or adjusted odds ratio (OR), relative risk (RR) or hazard ratio (HR) with 95% CIs  were reported; (3) studies were excluded if its published language is not English; (4) studies focused on a population with a disease such as renal disease and nonalcoholic fatty liver disease (NAFLD) were also excluded; (5) The publication with the largest sample size was considered eligible when there were multiple publications from the same study population. The CAD was defined as stable angina, unstable angina, myocardial infarction and sudden cardiac death. The diagnosis of CAD was based on the standards applied in all original articles. Studies with more than one subtype of CAD were regarded as two or more studies sharing the control group. Two investigators (HY Chen and JY Zhang) independently screened the titles and abstracts of records, and disagreements between investigators were resolved by consensus. Then, these investigators read full texts and confirmed studies for final inclusion with a third reviewer (ZQ Zhang).The necessary data were extracted independently by two researchers (HY Chen and JY Zhang) using a purpose-designed form. Any discrepancy was resolved by discussion or involved a third reviewer (ZQ Zhang) when necessary. Data were extracted from graphs using R package metaDigitise . The folThe risk of bias was assessed using the Risk of Bias In Non-Randomized Studies of Exposure (ROBINS-E) tool . The Grad and a summary OR with 95% CI were calculated to estimate the association of circulating RBP4 levels with CAD [I2 ≥ 30%); otherwise, a fixed-effects model was utilized as noted [2 or ≥ 25 kg/m2), complication status  and percentage of female subjects (≥50% or < 50%) to explore potential explanations for the heterogeneity. Meta-regression analyses were applied for exploring the possible sources of heterogeneity. The effect size was set as the dependent variable with geographic region (Asian or others), mean age of patients (< 60 years old or ≥ 60 years old), mean BMI of patients (< 25 kg/m2 or ≥ 25 kg/m2), female percentage among patients (≥50% or < 50%), quality score (continuous) and complication status  as explanatory variables. Sensitivity analyses were also performed by sequentially omitting each study to evaluate the robustness of the principal findings. Additionally, a cumulative meta-analysis was performed to monitor the evidence over time. Finally, publication bias was detected using Luis Furuya-Kanamori (LFK) index and the Doi plot [P values were two-sided, and a P value < 0.05 was considered statistically significant.The means and SD of circulating RBP4 levels in CAD and control groups and risk estimates were used to evaluate the effect size for each study. SMD presented as Cohen’s with CAD . RRs werwith CAD . A randoas noted . Howeveras noted . A qualias noted . In addias noted , 29. SubA flow chart of the study selection progress is presented in Fig. 2 across studies. Levels of RBP4 were determined using an ELISA in all of the studies. Most studies enrolled both female and male participants [n = 3169) vs controls (n = 3369); two studies, CAD with complications group (n = 177) vs controls (n = 141); and two studies, CAD alone and CAD with complications group (n = 125) vs controls (n = 130).Table icipants , 34–36, icipants , 31 and icipants , 33. TheThe results of the risk of bias assessment using the ROBINS-E tool are shown in Supplementary Table I2 = 96.00%, P < 0.01), the IVhet model and the random-effects model were adopted. As illustrated in Fig. As significant heterogeneity existed among studies  and complication status  might exert a significant effect on the pooled effect size, accounting for a low level of heterogeneity . In addition, geographic region , mean age of patients , female percentage  and quality score  failed to account for heterogeneity in the results  and used the IVhet and quality-effect models to show comparable circulating RBP4 levels in patients with CAD than in those without CAD. CAD patients with complications are particularly susceptible to having increased RBP4 levels.P<0.05) [A previous meta-analysis by Hou et al. including 10 studies with 1698 Chinese subjects reported a significant difference in RBP4 levels between CAD and non-CAD participants under the random-effect model  . The resP<0.05) . In addiP<0.05) . In contP<0.05) . TherefoA potential inflammatory mechanism was suggested to link circulating RBP4 levels to CAD, although the exact biological mechanism is currently unknown. Farjo et al. reported that RBP4 induces vascular inflammation by stimulating the mRNA expression of factors involved in leucocyte recruitment and adhesion to endothelial cells , 42. In P for the meta-regression analysis = 0.01). Patients with diabetes, NAFLD and other chronic diseases were commonly complicated with CAD [In the present meta-analysis, patients with CAD presenting complications  had a greater effect size than patients with CAD without complications compared with the controls  of PI was observed, indicating that RBP4 levels would be either significantly increased or decreased in future studies comparing patients with CAD and healthy controls. These results may reflect the uncertainty of the association between RBP4 levels and CAD risk, highlighting a limitation of epidemiological studies on this topic. Moreover, the risk of bias assessment revealed that most eligible studies had a moderate risk of bias, and the results obtained after adjustment for the risk of bias using the quality-effect model showed that the effects of the risk of bias on pooled effect sizes were limited in the current meta-analysis.The current meta-analysis has a much larger sample size and included literature from wider regions including China and other countries, compared with the previous meta-analysis . HoweverThere was low-quality evidence that circulating RBP4 levels in patients with CAD were comparable to the levels of controls. However, CAD patients with complications were at risk of having higher levels of RBP4. Based on the results, RBP4 may not be a risk factor for the progression of CAD. Comparisons among different subtypes of RBP4 with larger sample size are needed in the future.Additional file 1.


No previous meta‐analyses have compared the risk of dementia, due to an underlying atrial fibrillation (AF), in the short‐term versus the long‐term period.To perform an update meta‐analysis of studies examining the association between AF and dementia and the relative impact of follow‐up period.Data were obtained searching MEDLINE and Scopus for all investigations published between 1 January 2000 and March 1, 2021 reporting the risk of dementia in AF patients. The following MeSH terms were used for the search: “Atrial Fibrillation” AND “Dementia” OR “Alzheimer’s disease”. From each study, the adjusted hazard ratio (aHR) with the related 95% confidence interval (CI) was pooled using a random effect model.p < 0.0001; I2 = 93.5%) for dementia in subjects with AF. Stratifying the studies according to follow‐up duration, those having a follow‐up ≥10 years showed an aHR for dementia of 1.37 , while those with a follow‐up duration <10 years has a slightly higher aHR for dementia . Nine studies showed that the aHR for Alzheimer’s disease (AD) in AF patients was 1.30 .The analysis was carried out on 18 studies involving 3.559.349 subjects, of which 902.741 (25.3%) developed dementia during follow‐up. A random effect model revealed an aHR of 1.40 (95% CI: 1.27–1.54, Evidence suggests that patients with AF have an increased risk of developing dementia and AD. The risk of dementia was slightly higher when the follow‐up was shorter than 10 years. The presence of atrial fibrillation increased the risk of developing dementia by 40%The presence of atrial fibrillation increased the risk of developing Alzheimer’s disease by 40%The risk of developing dementia in patients with atrial fibrillation appears to be inversely related to the length of follow‐up The selection of studies was independently conducted by 2 authors  in a blinded fashion. Any discrepancies in study selection were resolved consulting a third author (CC). The following MeSH terms were used for the search: “Atrial Fibrillation” AND “Dementia” OR “Alzheimer’s disease.” Moreover, to ensure comprehensiveness, reference lists of retrieved studies and previous review articles were screened for additional relevant studies. Studies were included if: (1) they provided data regarding the risk of dementia in patients with confirmed AF; (2) the risk of dementia was expressed as adjusted hazard ratio (aHR) with relative 95% confidence (3) they reported information regarding the dementia diagnosis. Conversely, studies reporting the occurrence of AF in patients with mild cognitive impairment (MCI) as well as randomized controlled trials, case reports, review articles, abstracts, editorials/letters, and case series with less than 15 participants were excluded from the analysis. Pre‐clinical studies  were also excluded by the final analysis. If a study involved the same population, only the most recent investigation was included (overlapping cohort).2.2The primary outcome of the meta‐analysis was the development of any kind of dementia in AF patients. The secondary outcomes were the comparison of dementia risk after considering the duration of the follow‐up (<10 vs. ≥ 10 years) and the correlation between AF and risk of Alzheimer’s disease (AD). For all investigations reviewed we extracted the year of publication, gender , follow‐up duration, total number of participants and dementia patients and methods used for both AF and dementia diagnosis. Two authors (A.P. and C.B.) revised and extracted the data; in case of discrepancies a third author was consulted (C.C.).2.3The quality of included studies was graded using the Newcastle‐Ottawa quality assessment scale (NOS).2.42 statistic. Specifically, a I2 = 0 indicated no heterogeneity while we considered low, moderate, and high degrees of heterogeneity based on the values of I2 as <25%, 25%–75% and above 75% respectively. Moreover, tau‐squared (τ2) was also calculated to see the extent of variation among the effects observed in different studies. To evaluate the publication bias, both the visual inspection of the funnel plots and the Egger’s test were used. A predefined sensitivity analysis  was performed removing one study at the time. To further appraise the impact of potential baseline confounders, meta‐regression analyzed using the length of follow‐up of each study and the latitude of the population enrolled, as moderator variables were performed. Analyses were carried out using comprehensive meta‐analysis software (CMA), version 3. The HRs were compared by using the software R (R software—version 3.6.3) (http://www.r‐project.org/). A p‐value <0.05 was considered statistically significant.From each study, the adjusted hazard ratio (aHR) with the related 95% confidence interval (CI) was pooled using a random‐model while a forest plot was adopted to visually evaluate the results. Statistical heterogeneity between groups was measured using the Higgins I33.1A total of 1,460 articles were retrieved after excluding duplicates. The initial screening excluded 425 articles because they did not meet the inclusion criteria, leaving 674 articles to assess for eligibility. Subsequently, after evaluation of the full‐text articles, 656 were excluded and 18 investigations met the inclusion criteria  for dementia in AF subjects  revealed absence of bias. One‐by‐one exclusion of the studies from the analysis slightly changed the combined HR, which remained statistically significant across a range from 1.42  to 1.39 , suggesting that no single study had an undue impact on the combined HR. Meta‐regression analysis, using the length of follow‐up period of each study reviewed, showed a significant direct relationship with the risk of dementia due to AF .All 18 studies evaluated the relationship between AF and dementia risk. The variables used by each study reviewed for adjustments are presented in Table s Figure  and Egge3.3p < 0.0001, I2 = 96.6%; Figure p < 0.0001, I2 = 49%) . The relative funnel plots are showed in Figure The studies reviewed were further stratified according to the follow‐up duration to determine the risk of AF over time. Specifically, eight studies, based on 3.232.981 subjects, had a follow‐up ≥10 years ; in these patients, the aHR for dementia was 1.37 . Again, no bias was detected using the Egger’s test  or by the visual assessment using the funnel plot ; however, it is well‐known that the effect of a risk factor may vary during follow‐up. Two phenomena may have contributed to the trend we observed: (i) as for other risk factors  it is likely that the deleterious effect of AF may be reduced over the years (ii) since age is the strongest known risk factor for dementia it causes an important increase in the incidence of the disease also in subjects without AF, thus reducing the HR in the long‐term period.Our meta‐analysis confirms previous evidence showing an increased risk of dementia in patients affected by AF.Moreover, patients with AF had an additional 30% risk of developing AD compared to subjects without AF, and this confirms data from literature showing that AF is not only a risk factor for the form of dementia more directly linked with vascular dysfunction, that is vascular dementia (VD).The present results highlight that follow‐up period plays a pivotal role on the risk of dementia in AF patients. Indeed, recent investigations have elucidated the progressive nature of AF, that becomes more difficult to treat with increasing duration. This aspect is believed to be mainly due to the occurrence of electrical, contractile, and structural remodeling of the atria, which creates a fertile environment for the propagation of AF.Among the potential strategies suggested for the reduction of dementia onset due to an underlying AF, some investigations have suggested the potential benefices obtained from the administration of oral anticoagulants.Compared to the latest meta‐analysis available on this topic,Our analysis evidence how the preservation of cognitive performance and prevention of dementia should be fundamental goals in the management of elderly patients with AF. Besides the traditional cardiovascular evaluation, by using the CHADS2 and CHA2DS2‐VASc, to evaluate the risk of stroke and consequent need of anticoagulation therapy, a multidimensional geriatric evaluation should be recommended as an integral part of the management of these patients when suggestive symptoms for mild cognitive impairment or dementia are detected to limit the AF progression.The present study suffers from several limitations. Despite the use of a random‐effect model, a high heterogeneity was observed in the overall analysis, as well as after dividing the reviewed studies according to the duration of the follow‐up. Probably, this aspect depends on the inclusion criteria of participants, differences in sample size, methodological quality, demographic and ethno‐racial characteristics of the study populations, various covariate assessments, different length of follow‐up periods, types of dementia, and inherited bias from the original investigations. In particular, the different variables used for the statistical adjustments by each study may have contributed to the observed high heterogeneity. A further important limitation and source of heterogeneity of the meta‐analysis is the lack of information about the number of oral anti‐coagulant users (as well as the type of drugs) of the majority of the examined studies.5Patients with AF have an increased risk of developing dementia and AD, and the risk is slightly higher in the studies with a follow‐up shorter than 10 years. Further analyses are needed to confirm our results and to assess the potential benefit of a more aggressive therapy in those patients with AF and long‐life expectances. Indeed, elderly patients generally experience different comorbidities limiting the use of interventional treatments and therefore forcing to a medical treatment, which in several cases is not conclusive or allows some relapse of the arrhythmias with hemodynamic consequences and or complications. Therefore, it appears useful to adopt a patient’s tailored approach also considering the risk of dementia in the long‐term period which should promote the resolution of AF and potentially avoid watchful within approaches also if the arrhythmia is well tolerated and without hemodynamic effects.All authors declare that they have no conflicts of interest.Not necessary (systematic review).Supplementary Material 1Click here for additional data file.Supplementary Material 2Click here for additional data file.Supplementary Material 3Click here for additional data file.Supplementary Material 4Click here for additional data file.Supplementary Material 5Click here for additional data file.


Diabetes is one of the common chronic diseases in which susceptibility is determined by a combination of genetic and environmental factors, and more than 90% of diabetic patients are diabetes mellitus type 2 (T2DM). The existing studies on the association between CDKAL1 rs10946398 gene polymorphism and susceptibility to type 2 diabetes are inconsistent across populations.  We aim to explore the association between CDKAL1 rs10946398 gene polymorphism and susceptibility to type 2 diabetes in different populations. I2 was used to assess the heterogeneity ; correlation was judged by a forest map; potential publication bias was tested by the Egger test (p > 0.05 indicates that there is no publication bias).  We examined all studies before June 12, 2021, that associated CDKAL1 rs10946398 with T2DM. Heterogeneity was assessed by meta-analysis of allelic inheritance models (A vs. C), dominant inheritance models (AA vs. AC+CC), and recessive inheritance model (AA+AC vs. CC); p < 0.01) for dominant genetic model . But the allelic effect model  and the recessive genetic model  were not statistically significant (p > 0.01). In the non-Asian population, the differences were statistically significant (p < 0.01) for the allelic effect model , the dominant model , and the recessive model .  Fourteen data totaling 30288 subjects, including 19272 controls and 11016 patients with T2DM, met our inclusion criteria. In the Asian population, the differences were statistically significant ( In this study, CDKAL1 RS10946398 was positively associated with T2DM, but the association was different in Asian populations. According to the World Health Organization (WHO), approximately 3.4 million people died from developing diabetes in 2004, and it predicts that the number of diabetes deaths will double between 2005 and 2030. The International Diabetes Federation predicts that the global prevalence of diabetes will reach 642 million cases by 2040 , with type 2 diabetes accounting for more than 90% of diabetics [Type 2 diabetes mellitus (T2DM), formerly known as non-insulin-dependent or adult-onset diabetes mellitus, is a type of diabetes mellitus. It is caused by poor insulin action which is the relative lack of insulin in patients, and its susceptibility is determined by both genetic and environmental factors [Previous studies have shown that China –6 and otAn advanced search of the literature search library was conducted by using “T2DM CDKAL1” and “CDKAL1 rs10946398” as key to search terms in the China National Knowledge Infrastructure (CNKI), PubMed, and WanFang digital databases, with the last search conducted on June 12, 2021.Case-control studies focus on the association between the CDKAL1 rs10946398 polymorphism and T2DM in adultsPatients were randomly selected with no special restrictions on gender, family history, etc.Studies provide accurate control and case group data sourcesThe data provided in the study report were statistically significant. The study results had specific OR values, 95% CIStudies met the diagnostic criteria of T2DM published by the World Health Organization (WHO) in 2019, and the control group all met the law of H-W genetic balanceThe following studies were included , 17:CasThere were only case groups or a lack of sufficient controlsStatistical data are erroneous or there are significant differences in the statistics of the same study in different literaturesThe overall sample size is insufficientLiterature reviews and case reports were excludedThe following studies were excluded:Two investigators independently read the literature and extracted information from the eligible literature based on exclusion and inclusion criteria. In case of ambiguity, a consensus was reached on whether to extract the paper data through discussion with the third investigator. For each paper, the following information was collected: (i) author's name, (ii) year of publication, (iii) ethnicity and country of the study population, (vi) number of included cases and controls, and (vii) genotype data . The litReview Manager 5 software was used to complete the meta-analysis. Stata software was used to complete the Egger test.We obtained articles on the relationship between CDKAL1 rs10946398 diversity and T2DM susceptibility from PubMed and CNQ. After reading the title, year, author, and abstract of the papers, we conducted the first screening. The second screening was performed by reading the full text and analyzing whether the data was statistically significant. Finally, 14 literatures were included. A total of 14 datasets were obtained for meta-analysis by reading through the full text to filter the data required for recording. A total of 30288 subjects were included in the meta-analysis, including a total of 11016 in the T2DM patient group and 19272 in the control group. Eight of the datasets were from the Asian study population: 3 from China, 1 from India, 1 from Korea, 1 from Japan, 1 from Iran, and 1 from the United Arab Emirates; 6 were from the non-Asian study population: 3 from the USA, 2 from Russia, and 1 from Mexico. Information on the first author, study year, sample size, ethnicity, BMI, mean age of control and case groups, and risk allele frequency for each study is shown in In the evaluation of the relationship between the CDKAL1 rs10946398 gene and T2DM susceptibility, a total of 14 studies were included in the meta-analysis after literature data search, screening, and verification. In order to analyze the association between CDKAL1 rs10946398 polymorphism and susceptibility to T2DM, we analyzed the relationships between A and C alleles, AA+AC and CC genotypes, AA and AC+CC genotypes in T2DM patients and controls. Since 8 studies were from Asia and 6 were from non-Asia, we stratified the Asian and non-Asian populations.I2, to assess the magnitude of heterogeneity . Because our data were randomly selected and we wanted to reflect the overall situation with a small sample size, only the allelic genetic model and the recessive genetic model in non-Asian populations show that I2 < 50%, so we used the random-effects model  for the allelic genetic models , the dominant genetic models , and the recessive genetic models . The results are shown in Figures 2In the total population, the differences were statistically significant (p < 0.01) for dominant genetic model . But the allelic effect model  and the recessive genetic model  were not statistically significant (p > 0.01). The results are shown in Figures 2In the Asian population, the differences were statistically significant (p < 0.01) for the allelic genetic model , the dominant genetic model , and the recessive genetic model . The results are shown in Figures 2In non-Asian populations, the differences were statistically significant , recessive inheritance model (AA+AC vs. CC), and dominant inheritance models (AA vs. AC+CC) were 0.114, 0.307, and 0.304, respectively, which were greater than 0.05, indicating that there was no publication bias. What is more, according to the symmetry of the funnel plot, the existence of publication bias can also be judged. The results are shown in Figures 5We used Stata software for the Egger test, and the β cells continue to be activated, which may inhibit the activity of CDK5 in pancreatic β cells. Insulin secretion is reduced by lowering the expression of insulin genes [According to a large number of genome-wide association analyses (GWAS), CDK5 regulation-related protein 1-LIAK 1 (CDKAL1) gene under the action of high glucose toxicity will increase the body's demand for insulin, and pancreatic in genes , 19. Becin genes . In partin genes .β cell function. CDKAL1rs10946398 also plays an important role in the pathogenesis of T2DM in the detected Russian population [To study the relationship between the variation of CDKAL1 RS10946398 locus and the susceptibility to T2DM in different populations, 14 sets of data were finally used for meta-analysis through data investigation and screening, and 13 sets of data showed that the CDKAL1 RS10946398 locus was significantly correlated with the incidence of T2DM; for example, a study by Nfor et al. showed a significant association between CDKAL1 RS10946398 and T2DM in Taiwanese. CC carriers were more associated with T2DM than AC carriers, and C allele carriers were more associated with type 2 diabetes than A allele carriers [pulation . Only onIn this study, a meta-analysis of the included 14 groups of data concerning the CDKAL1 rs10946398 locus and T2DM was performed by analyzing allelic models (A vs. C), recessive genetic models (AA+AC vs. CC), and dominant genetic models (AA vs. AC+CC) in T2DM patients and controls. Of the 30288 subjects, including 19272 controls and 11016 T2DM patients, we found that CDKAL1 RS10946398 gene polymorphism locus is associated with type 2 diabetes mellitus in different ethnic groups, and the degree of correlation is different in different genetic models.p < 0.01) for the dominant genetic model . But the allelic effect model  and the recessive genetic model  were not statistically significant (p > 0.01). The risk ratio of the A allele was higher than that of the C allele. In the non-Asian population, the differences were statistically significant (p < 0.01) for the allelic effect model , the dominant model , and the recessive model . The risk ratio of the A allele was higher than that of the C allele [In the Asian population, the differences were statistically significant , other models were statistically significant (p < 0.01). CDKAL1rs10946398 could significantly increase the risk of T2DM in the allele model of Asian and in all models of non-Asian. But this result cannot be attributed to differences in ethnicity; it could also be due to the small sample size. In conclusion, the CDKAL1 rs10946398 gene variant may increase the susceptibility to T2DM.We used 14 sets of data for meta-analysis of the locus genetic model (A vs. C) recessive models (AA+AC vs. CC) and dominant models (AA vs. AC+CC). Except for the Asian allelic effect model and recessive gene model (