What are the quality and certainty of evidence for an association between hormonal contraceptive use and the risk of adverse health outcomes in meta-analyses of randomized clinical trials and cohort studies?In this umbrella review of 58 meta-analyses of randomized clinical trials and cohort studies describing 156 associations between hormonal contraceptive use and adverse health outcomes among women, no associations with adverse outcomes, including cardiovascular and cancer risk, were supported by high-quality evidence. However, the association between the use of a levonorgestrel-releasing intrauterine system and reductions in endometrial polyps associated with tamoxifen use was graded as having high-quality evidence.The results of this umbrella review supported preexisting understandings of the risks and benefits associated with hormonal contraceptive use, suggesting that the associations between hormonal contraceptive use and adverse health outcomes are not supported by high-quality evidence. Meta-analyses have reported conflicting data on the safety of hormonal contraception, but the quality of evidence for the associations between hormonal contraceptive use and adverse health outcomes has not been quantified in aggregate.To grade the evidence from meta-analyses of randomized clinical trials (RCTs) and cohort studies that assessed the associations between hormonal contraceptive use and adverse health outcomes among women.hormonal contraception, contraceptive agents, progesterone, desogestrel, norethindrone, megestrol, algestone, norprogesterones, and levonorgestrel combined with terms such as systematic review or meta-analysis.MEDLINE, Embase, and the Cochrane Database of Systematic Reviews were searched from database inception to August 2020. Search terms included The methodological quality of each meta-analysis was graded using the Assessment of Multiple Systematic Reviews, version 2, which rated quality as critically low, low, moderate, or high. The Grading of Recommendation, Assessment, Development and Evaluations approach was used to assess the certainty of evidence in meta-analyses of RCTs, with evidence graded as very low, low, moderate, or high. Evidence of associations from meta-analyses of cohort studies was ranked according to established criteria as nonsignificant, weak, suggestive, highly suggestive, or convincing.P ≤ .05); no associations between hormonal contraceptive use and adverse outcomes were supported by high-quality evidence. The association between the use of a levonorgestrel-releasing intrauterine system and reductions in endometrial polyps associated with tamoxifen use  was graded as having high-quality evidence, and this evidence ranking was retained in the subgroup analysis. Among meta-analyses of cohort studies, 40 of the 96 associations were nominally statistically significant; however, no associations between hormonal contraceptive use and adverse outcomes were supported by convincing evidence in the primary and subgroup analyses. The risk of venous thromboembolism among those using vs not using oral contraception  was initially supported by highly suggestive evidence, but this evidence was downgraded to weak in the sensitivity analysis.A total of 2996 records were screened; of those, 310 full-text articles were assessed for eligibility, and 58 articles  were selected for evidence synthesis. Sixty associations were described in meta-analyses of RCTs, and 96 associations were described in meta-analyses of cohort studies. Among meta-analyses of RCTs, 14 of the 60 associations were nominally statistically significant (The results of this umbrella review supported preexisting understandings of the risks and benefits associated with hormonal contraceptive use. Overall, the associations between hormonal contraceptive use and cardiovascular risk, cancer risk, and other major adverse health outcomes were not supported by high-quality evidence. This umbrella review assesses evidence from meta-analyses of randomized clinical trials and cohort studies examining the association between hormonal contraceptive use and adverse health outcomes among women. These drugs are available in different dosages and forms, such as tablets, implants, intrauterine systems, intramuscular injections, intravaginal rings, and skin patches.2 Overall, when used correctly and consistently, all hormonal contraceptive agents  have been found to be effective, safe, and reversible forms of contraception.3Hormonal contraceptive use has been increasing worldwide. In 2019, approximately 1.1 billion women of reproductive age were in need of family planning services worldwide, with almost 50% of those women using various hormonal contraceptive methods,4 on the associations between hormonal contraceptive use and adverse health outcomes. The use of hormonal contraception has been associated with either reduced or increased risk of many adverse health outcomes, including cancer, cardiovascular, fracture, gastrointestinal, and metabolic outcomes.4 These risks vary depending on the dose, route, duration of use, and generations and formulations of hormonal contraceptive agents. Because this heterogeneity is present in previous meta-analyses, robust grading of the evidence is needed. To our knowledge, there has been little synthesis of the certainty and quality of the evidence in aggregate across meta-analyses. Umbrella reviews summarize the evidence from multiple meta-analyses on the same topic by assessing the certainty and precision of the associations and the presence of bias, thereby enabling the grading of evidence using well-defined criteria.5However, contradictory results have been reported from several meta-analysesThis umbrella review aimed to systematically identify relevant meta-analyses of randomized clinical trials (RCTs) and cohort studies of hormonal contraceptive agents, summarize their findings, and assess the certainty of their evidence to provide a comprehensive understanding of the associations between hormonal contraception and adverse health outcomes.CRD42021231959). The study followed the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) reporting guideline6 and the Meta-analysis of Observational Studies in Epidemiology (MOOSE) reporting guideline.7The protocol for this study was registered at the International Prospective Register of Systematic Reviews . We did not include meta-analyses of nonspecific adverse events, such as abdominal pain and similar gastrointestinal discomforts, acne, leucorrhea, or breast changes. Adverse health outcomes reported in meta-analyses were extracted as defined by the original authors of each meta-analysis. When more than 1 meta-analysis was available for the same research question, we selected the meta-analysis with the largest data set, as previously described elsewhere.Titles and abstracts of articles were screened by 2 reviewers (S.B. and M.K.) independently. The same 2 reviewers independently assessed the full text of potentially eligible articles. Any discrepancies were resolved by a third reviewer (S.V.) via consensus.11Data extraction was independently performed by 2 reviewers (S.B. and V.R.), with any disagreement referred to a third reviewer (S.V.) for resolution. We extracted data at both the meta-analysis and individual study levels  were graded as having convincing (class 1), highly suggestive (class 2), suggestive (class 3), or weak (class 4) evidence based on the amount of evidence, statistical significance, heterogeneity, small-study effect, excess significance bias, prediction interval, and credibility ceiling test  framework, which classified evidence as very low, low, moderate, or high.14 Another sensitivity analysis was conducted based on the approach used for random-effects meta-analyses. We used the Hartung-Knapp-Sidik-Jonkman approach rather than the DerSimonian and Laird method for associations reported in fewer than 5 studies.15We performed a series of sensitivity analyses to determine the robustness of the findings for each association that was initially graded as having high or moderate evidence (for RCTs) or convincing or highly suggestive evidence (for cohort studies). These sensitivity analyses were performed by excluding small studies (<25th percentile) and primary studies with high risk of bias or low-quality evidence.16 We assessed heterogeneity using the I2 statistic.17 The evidence for small-study effects was evaluated using the Egger test.18For each association between hormonal contraceptive exposure and outcome, we extracted effect sizes of individual studies included in each meta-analysis based on study design, and we repeated the meta-analyses separately for RCTs and cohort studies to calculate the pooled effect sizes and 95% CIs using random-effects models.19 We also applied excess significance assessment using an χ2 test, which evaluated whether the observed number of studies with statistically significant results differed from the expected number of studies with statistically significant results.21 We also performed a credibility ceiling test using a ceiling of 10% to reflect our level of confidence in the cohort studies.23 All statistical analyses were conducted using Stata software, version 16.0 (StataCorp LLC). The level of significance was set at 2-sided P = .05 for all tests, with the exception of the Egger and excess significance tests, which had significance levels of 2-sided P = .10.For meta-analyses of cohort studies, we also estimated the 95% prediction interval, which evaluated the uncertainty of effect that would be expected in a new population addressing a similar association.81 were selected for evidence synthesis; of those, 13 articles81 were meta-analyses of RCTs, and 45 articles79 were meta-analyses of cohort studies. The list of articles excluded after the selection criteria for the overlapping meta-analyses were applied is provided in eTable 3 in the In total, we screened 2996 records and assessed 310 full-text articles for eligibility. Of those, 102 articles (32.8%) were eligible for preliminary data extraction , with a median sample of 91 participants  and a follow-up duration ranging from 3 months to 10 years. Additional descriptive characteristics are provided in eTable 4 in the Thirteen eligible meta-analyses81 using the AMSTAR-2 tool revealed that 8 meta-analyses74 (61.5%) were of high quality, 3 meta-analyses72 (23.1%) were of low quality, and 2 meta-analyses81 (15.4%) were of critically low quality  were nominally statistically significant (P ≤ .05) based on random-effects models. A total of 21 associations35 (35.0%) had high heterogeneity (I2 > 50%). Small-study effects were found for 3 associations26 (5.0%). Summaries of all significant and nonsignificant associations are provided in Fourteen27 7 associations26 reported an increased risk of adverse health outcomes but no major adverse health outcomes   compared with women who were not using this system. Another association25 with moderate-quality evidence reported an increased fasting insulin level with combined oral contraceptive use  for the treatment of PCOS . No sensitivity analyses were conducted for these 2 associations25 because of the limited number of studies.Among the 14 statistically significant associations,utcomes) . For tre27 between hormonal contraceptive use and reductions in the risk of adverse health outcomes . A subgroup analysis of this association including only postmenopausal women retained the same evidence ranking  and reductions in fasting blood glucose (FBG) levels among women with PCOS was graded as having moderate-quality evidence . This association was upgraded to high-quality evidence after a sensitivity analysis excluding studies with high risk of bias . In addition, women with PCOS experienced increases in low-density lipoprotein levels  and total cholesterol levels . Associations that were initially graded as having evidence of very low to high quality in the primary analysis are provided in eFigure 6 and eFigure 7 in the We identified 7 associationsoutcomes . One ass79 described 96 potential associations, including 171 individual study estimates of adverse health outcomes associated with hormonal contraceptive exposure. The median number of studies per meta-analysis was 3 , and the follow-up duration ranged from 3 months to 42 years. The median sample based on 47 associations was 199 participants . For 29 of 96 associations (30.2%), the number of participants was greater than 1000. Additional descriptive characteristics are provided in Forty-five eligible meta-analyses79 using the AMSTAR-2 tool revealed that 6 meta-analyses77 (13.3%) were of high quality, 14 meta-analyses78 (31.1%) were of moderate quality, 7 meta-analyses79 (15.6%) were of low quality, and the remaining 18 meta-analyses71 (40.0%) were of critically low quality  were nominally statistically significant (P ≤ .05), and only 4 associations78 (4.2%) reached high significance with a P value of 1.00 × 10−6 or less (78 (39.6%) had high heterogeneity (I2 > 50%). The 95% prediction interval excluded the null value for only 5 associations78 (5.2%). Small-study effects were found for 10 associations69 (10.4%), and no excess significance bias was observed for any association. Among 40 statistically significant associations,78 only 18 associations78 (45.0%) passed the credibility ceiling test using a ceiling value of 10%. Summaries of all significant and nonsignificant associations are provided in Forty or less . Thirty-78 with an increased risk of adverse health outcomes, including cardiovascular risk, cancer risk, or other major adverse outcomes , were statistically significant  was supported by class 1 evidence . Two associations65 revealing increased risk of venous thromboembolism (VTE) were graded as having class 2 evidence. One of those associations65 was among women using oral contraception compared with women not using oral contraception , whereas the other association44 was among women using low-dose combined oral contraception  compared with women using a levonorgestrel-releasing intrauterine system . Among women with PCOS receiving combined oral contraception, findings suggested increased triglyceride levels . In addition, women with PCOS experienced an increase in total cholesterol levels .Thirty associationsnificant . One of 46 was downgraded from class 1 to class 4 after a subgroup analysis excluding the population with surgically diagnosed endometriosis  and age, 2 factors known to be associated with VTE risk. Overall, this small variance was unlikely to be clinically meaningful when selecting a type of combined oral contraception.All meta-analyses of cohort studies evaluating estrogen-containing hormonal contraceptive agents25 Therefore, it was difficult to extrapolate these associations to the general population because women with PCOS are typically obese and at risk of metabolic syndrome, insulin resistance, and hyperlipidemia. Estrogen modifies lipid metabolism in the liver and regulates serum lipoprotein levels, resulting in increased high-density lipoprotein and triglyceride levels and decreased low-density lipoprotein and total cholesterol levels.83 Progestin type can also have consequences for the extent of lipid profile change induced by exogenous estrogen.84 Notably, the types of triglyceride particles increased by ethinyl estradiol have not been associated with increases in the risk of atherosclerotic disease.85 Among women with PCOS using combined oral contraception, the findings from our review of meta-analyses of RCTs suggested increases in high-density lipoprotein levels (mean difference ranging from 6.50 to10.0025), and the findings from our review of meta-analyses of cohort studies suggested increases in triglyceride levels (mean difference ranging from 0.7351 to 39.8225). Contrary to previous evidence,86 our review found an increase in low-density lipoprotein levels  and total cholesterol levels  among women with PCOS. This difference potentially represents the consequences of different progestin formulations, follow-up durations, and compounding of factors underlying the metabolic changes of PCOS. The quality of evidence was rated class 4 in the cohort studies47 and ranged from low to very low in the RCTs25 for associations between women using combined oral contraception and lipid changes, such as increased total cholesterol and triglyceride levels. In addition, the absolute increase in laboratory values was unlikely to be clinically meaningful.In our umbrella review, metabolic changes were only assessed among women with PCOS.25 had moderate- to high-quality evidence for an association between hormonal contraceptive use and increases in fasting insulin levels as well as decreases in FBG levels among women with PCOS. Although these findings appear to be contradictory, the progestin formulations in the tablets were different; desogestrel was used in the RCT reporting increased fasting insulin levels, and cyproterone acetate was used in the RCT reporting decreased FBG levels. The consequences of combined oral contraception for insulin sensitivity were mainly associated with the progestin component; therefore, the antiandrogenic properties of cyproterone acetate may be associated with decreases in insulin resistance and reductions in circulating androgen levels among women with PCOS, thereby producing the FBG levels observed.87 However, the clinical importance of the impact of different progestins was likely to be minimal, and the use of any combined oral contraceptive agent was unlikely to be associated with diabetes among women with PCOS.88One meta-analysis of RCTs78 we found evidence of an approximately 1.5-fold increase in the risk of breast cancer54 and an approximately 1.5- to 2.5-fold increase in the risk of cervical cancer78 with the use of combined oral contraception, although the evidence was of class 3 and class 4 quality. Notably, the breast cancer associations reported54 were among women with BRCA1 (OMIM 113705) and BRCA2 (OMIM 600185) variants rather than women in the general population. Individuals with BRCA1 or BRCA2 variants have an increased baseline risk of breast cancer, and the additive impact of combined oral contraceptive risks would be magnified.90 Current and recent use of estrogen-containing hormonal contraception is thought to increase breast cancer risk through a tumor promoter effect rather than an initiator effect on preexisting cancer cells.92 In addition, any increased risk of breast cancer returns to baseline 10 years after cessation of combined oral contraception.93In 2 meta-analyses of cohort studies,78 Few studies included in that meta-analysis controlled for human papilloma virus infection status, which is an important factor associated with cervical cancer risk. In the meta-analysis by Asthana et al,78 the subgroup analysis of studies including participants with known human papilloma virus infection suggested a modest increase in risk associated with combined oral contraceptive use. However, women using combined oral contraception have been reported to have more sexual partners and higher rates of human papilloma virus infection and to receive a greater number of Papanicolaou tests.94 A combination of these factors likely explains the increased risk of cervical cancer among women currently using combined oral contraception. Similar to breast cancer risk, the risk of cervical cancer appears to be time sensitive, with little to no increased risk when combined oral contraception has been used for less than 5 years or has not been used for more than 10 years.95This umbrella review also found class 4 evidence to suggest an association between combined oral contraceptive use and an increased risk of cervical cancer in a meta-analysis by Asthana et al.27 and decreases in the incidence of ovarian and colorectal cancers.64 The most significant association that was supported by high-quality evidence was found in a meta-analysis of RCTs27 in which the use vs nonuse of a levonorgestrel-releasing intrauterine system was associated with a 78% reduction in the risk of endometrial polyps among those receiving tamoxifen. Tamoxifen, a nonsteroidal selective estrogen receptor modulator, is known to induce the formation of endometrial polyps through its estrogen agonist effects on the endometrial lining.96 The use of a levonorgestrel-releasing intrauterine system mitigates this risk through its suppressive action on endometrial proliferation.97 Of note, controversy exists regarding the use of a levonorgestrel-releasing intrauterine system among patients with progesterone receptor–positive breast cancer because of the potential risk of disease recurrence.98 Additional noncontraceptive benefits associated with combined oral contraception include reductions in the risk of ovarian52 and colorectal cancers,64 and the results of our review supported these findings. Although these associations67 were graded as class 4 in quality, they have also been observed in multiple epidemiological studies.101Our umbrella review suggested several benefits associated with hormonal contraception, including reductions in endometrial polyps in the setting of tamoxifen use102 and the Centers for Disease Control and Prevention104 can be used to select appropriate hormonal contraception for specific groups of patients based on these risk factors. Future research would ideally include well-designed RCTs comparing adverse health outcomes among those using vs not using hormonal contraception. However, the feasibility of conducting RCTs large enough to detect many of these adverse outcomes is limited. Rigorous cohort studies may aid in better delineating these risks.87Robust grading of previous meta-analyses offers clinicians a tool to better evaluate hormonal contraceptive recommendations for at-risk patients. This grading may assist health care practitioners in the selection of hormonal contraceptive agents for their patients and encourage the consideration of existing individual risk factors, such as VTE, cancer, and cardiovascular risk. The medical eligibility criteria guidelines published by the World Health OrganizationThis study has several limitations. The umbrella review focused on existing meta-analyses. We found that some adverse outcomes were not included in these meta-analyses, precluding us from performing a comprehensive evaluation of safety aspects of hormonal contraceptive agents. The quality of all primary studies included in each meta-analysis relied on the assessment reported by the respective meta-analysis. Most systematic reviews and meta-analyses included in our review were focused on estrogen-containing combined oral contraceptive agents. When drawing conclusions about clinical practice from the findings of this umbrella review, it is necessary to be mindful that progestin-only methods  are not represented in a clinically meaningful way. However, the conclusions drawn from the combined oral contraceptive data do have implications for clinical practice.In this umbrella review, the associations between hormonal contraceptive use and cardiovascular risk, cancer risk, and other major adverse health outcomes were not supported by high-quality evidence. The findings reinforced preexisting understandings of the risks and benefits associated with hormonal contraceptive agents.


Background & Aims: Nonalcoholic fatty liver disease (NAFLD) is the most common liver injury. We performed this umbrella review of meta-analyses to summarize the evidence on the associations of nutritional, lifestyle, and metabolic factors with NAFLD.Methods: We searched the PubMed, Embase, and Web of Science databases from inception until July 2, 2020, to identify meta-analyses of observational studies which explored the associations of nutritional, lifestyle, and metabolic factors with NAFLD. Evidence levels were assessed using summary effect sizes, 95% prediction intervals, between-study heterogeneity, evidence of small-study effects, and evidence of excess significance bias for each meta-analysis. .Results: Twenty two risk or protective factors from 10 published meta-analyses were included and studied. Three risk factors  with highly suggestive levels of evidence and three risk factors  with suggestive levels of evidence were identified. Only two protective factors  with suggestive levels of evidence were identified. Furthermore, other six risk factors and two protective factors with weak levels of evidence were identified.Conclusions: We found varying levels of evidence of associations of nutritional, lifestyle, and metabolic factors and NAFLD. The results suggest that nutritional and lifestyle management should be considered as a major primary preventive strategy for NAFLD. Moreover, considering the low quality of included meta-analyses and limited area of research topics, future high-quality original studies and meta-analyses should be performed to study these associations. Non-alcoholic fatty liver disease (NAFLD) has variable disease courses, including simple steatosis, non-alcoholic steatohepatitis, and cirrhosis . NAFLD nA growing body of evidence suggested that both genetic and environmental factors  contribute to the development of NAFLD –11. MoreTo the best of our knowledge, no umbrella review has systematically concluded on the overall associations between diverse nutritional, lifestyle, and metabolic factors (risk or protective) and NAFLD. To fill this gap in the literature, we performed this umbrella review to summarize the results of existing meta-analyses that have explored the associations between nutritional, lifestyle, and metabolic factors and NAFLD in both adults and children. Further, we assessed the robustness of the evidence to inform future preventive strategies.We followed a standardized methodology and reported the findings of this umbrella review according to the recommendations of the Meta-analyses of Observational Studies in Epidemiology recommendations . The proTo identify systematic reviews or meta-analyses of observational studies that examined the associations of a number of nutritional, lifestyle, and metabolic factors with NAFLD, we systematically and comprehensively searched the PubMed, Embase, and Web of Science databases from their inception to July 2, 2020, without restrictions. The search keywords of each database are provided in Publications were initially screened based on the title and abstract. The full texts of potentially eligible publications were scrutinized independently by two investigators (H-XD and Y-SL). If these two investigators disagreed, a third investigator (Q-JW) were applied to make the final decision. Studies were included if they met the following criteria: (1) meta-analyses of any one of the observational study designs ; (2) the exposure of interest as nutritional, lifestyle, and metabolic factors; (3) reporting on the NAFLD as the outcome; (4) reporting of the usable risk estimates  between nutritional, lifestyle, and metabolic factors and NAFLD; (5) published in the English; (6) including both adults and children. Articles that described separate meta-analyses of more than one nutritional, lifestyle, and metabolic factors were included separately. Furthermore, for more than one meta-analysis on the same association, we included the one which assessed the largest number of primary studies. For meta-analyses on the same association that included the same number of primary studies, we included the one with the largest amount of prospective data.We excluded meta-analyses that investigated the association between genetic markers and the risk of NAFLD. Trials were unavailable for our research question. We excluded systematic reviews that did not feature quantitative analyses, meta-analyses based on individual data without systematic review, or articles that included animal trials or laboratory studies. We also excluded systematic reviews or meta-analyses that lacked study-specific data . In addition, studies that examined NAFLD as a risk factor for other medical conditions or diseases were also excluded.For each eligible meta-analysis, we extracted the name of the first author, journal name, publication year, study design, number of studies included, study population, nutritional, lifestyle, and metabolic factors, outcome(s) of interest investigated, and type of effect metric. Moreover, we extracted information on each primary study in the included meta-analyses, including the exposure, first author, publication year, the number of cases and controls (case-control studies), participants, the length of follow-up (cohort studies), risk estimates, and 95% CIs. We also extracted information relating to dose-response relationships from all meta-analyses. We included studies reporting different measures of effect size (ES) . Missing measures were calculated if sufficient data were available. Two independent investigators (J-LL and HS) extracted data from eligible publications. Discrepancies between the data extracted by the two investigators were resolved by a third investigator (Q-JW).The investigators (Y-TJ and YX) independently assessed the methodological quality of qualified systematic reviews and meta-analyses using AMSTAR 2 (A Measurement Tool to Assess systematic Reviews) . DiscrepP values, using both fixed- and random-effects models  for the random-effects estimates were calculated to account for between-study heterogeneity and to represent the possible range in which the risk estimates of new studies might lie , if the 95% PIs excluded the null hypothesis, if the I2 statistic for heterogeneity was <50%, if there was no evidence of small study effects (P > 0.10), and if excess significance bias (P > 0.10) was indicated.The evidence was defined as convincing when the P value for the random-effects model was <10−6, if the meta-analysis included more than 1,000 cases or more than 20,000 participants for continuous outcomes, and if the largest component study reported a significant result.The evidence was defined as highly suggestive if the P value for random-effects was <10−3, or if there were more than 1,000 cases or more than 20,000 participants for continuous outcomes.The evidence was defined as suggestive if the P value for significant associations was <0.05.The evidence was defined as weak if the P value > 0.05.We used the ‘non-significant associations' classification if all association tests yielded a n = 3, 25%) or critically low .The results of the systematic search and selection of eligible studies are presented in Four out of 11 dietary factors were found to be associated with NAFLD , 2. IntaAs shown in We explored the associations between the levels of four blood biomarkers and NAFLD , 2. SeruWaist circumference, body mass index (BMI), and waist-to-hip ratio were all determined as metabolic risk factors for NAFLD , 2. For To the best of our knowledge, our study is the first umbrella review that provides a comprehensive overview and critical assessment of risk and protective factors for NAFLD. A total of 22 risk or protective factors  from 10 published meta-analyses were included and studied. Regarding risk factors, three factors  with highly suggestive levels of evidence and three factors  with suggestive levels of evidence were identified. Only two protective factors  with suggestive levels of evidence were identified. Furthermore, six risk factors and two protective factors with weak levels of evidence were also identified. This umbrella review summarized the results based on existing meta-analyses. However, many novel and latest factors are not included in this study because of few meta-analyses of these topic. For example, pervious study found that seaweed consumption . intake We found that 66% of the examined associations had large heterogeneity (>50%) and 48% of them suffered from small-study effects or/and excess significance. Although these features could have resulted from genuine differences across the studies included, the results also suggested that biased results existed in some of the original studies included and subsequently caused concerns in the epidemiological credibility of these meta-analyses. Thus, in the present umbrella review, in accordance with previously published umbrella reviews –27, we cP value of the random-effects model for the association between waist circumference and NAFLD also contributed to the results. Moreover, considering that visceral obesity is a risk factor for a number of complications of metabolic syndrome, assessment of waist circumference may be more sensitive than BMI (We found that waist circumference (highly suggestive) and BMI (suggestive) were both risk factors for NAFLD. Obesity, particularly visceral obesity, has long been established as a major risk factor for NAFLD . Viscerathan BMI . Howeverthan BMI . Thus, ithan BMI . Indeed,than BMI . In termde novo lipogenesis, blocking β-fatty acid oxidation, and improving insulin resistance  and soft drinks (suggestive) and NAFLD. The consumption of fructose, which is the major sweetener in sugar-sweetened beverages and soft drinks, was strongly suggested as a major dietary risk factor for NAFLD . Firstlysistance , 52, 53.sistance , 55. Altsistance ] seems it drinks . There iBoth healthy diet and increased physical activities were considered as major lifestyle interventions for the management of NAFLD according to most current guidelines . We founMoreover, we found that former smoking was a lifestyle risk factor for NAFLD based on suggestive evidence. A possible explanation for this association could be that smoking cessation could induce weight gain . A previFinally, among blood biomarkers, we identified that serum fetuin-A level was a risk factor for NAFLD based on highly suggestive evidence while vitamin D level was a protective factor for NAFLD based on suggestive evidence. Serum fetuin-A, which is secreted by both the liver and adipose tissue, can contribute to the development of NAFLD by increasing the secretion of inflammatory cytokines via macrophages , 69. MorOur study had several limitations. Firstly, we only considered associations between risk/protective factors and NAFLD that have been assessed in meta-analyses of observational studies. Studies that have not been evaluated through meta-analyses were missed. Secondly, considering that many cross-sectional studies were included in the meta-analyses, it was impossible to infer causality. However, few meta-analyses that explored the associations between nutritional, lifestyle, and metabolic factors and NAFLD included cohort studies and/or randomized controlled trials only. More cohort studies and randomized controlled trials are needed to provide more robust pieces of evidence of the associations between risk/protective factors and NAFLD. Thirdly, AMSTAR 2 was used to evaluate the methodologic quality of the included meta-analyses. However, the qualities of the meta-analyses included were low . Most of the meta-analyses included were not included in the pre-registered protocol. All the meta-analyses included did not explain their selection of the original study designs for inclusion, provide a list of original studies excluded and justify the exclusions, or report on the sources of funding for the original studies included in their meta-analyses. However, the qualities of the included meta-analyses according to AMSTAR 2 would not affect the evidence grading for the associations between nutritional, lifestyle, and metabolic factors and NAFLD in the present umbrella review. Thus, future high-quality meta-analyses that adhere to the principles of AMSTAR 2 are needed. In addition, we found that all the included meta-analyses were low-quality. It may be due to the fact that the 16 items in the AMSTAR2 involved meticulous information, and some may not be presented in the article. Therefore, we try to contact the corresponding authors to request the details information of methods. However, we did not receive response from the authors. Fourthly, the high vs. low analysis is relatively indefinite and will lose much information in the initial study. For this reason, the guiding significance for the population to improve NAFLD is relatively limited. This bias can be avoided by dose-response analysis, but only eight dose-response meta-analysis were included in this study. Therefore, this result should be interpreted with caution. In conclusion, future dose-response initial studies are needed to interpret the summary estimates of meta-analysis and umbrella review more reasonably. Furthermore, we considered as eligible studies only published studies and did not include unpublished data. Thus, the results could be affected due to data loss. Finally, umbrella review is a method that synthesizes a large number of existing meta-analyses, rather than performing the included meta-analyses from the beginning . Thus, wIn summary, our study assessed the evidence on 22 associations between nutritional, lifestyle, and metabolic factors and NAFLD. Without considering genetic influences, generally, unhealthy lifestyle  occurred first and induced subsequent metabolic change (such as obesity). In this process, several anthropometric characteristics and blood biomarkers were identified, such as waist circumference, BMI, serum fetuin-A level, and vitamin D level. The results suggested that lifestyle management should be considered as a major primary preventive strategy for NAFLD. Moreover, considering the low quality of meta-analyses included (assessed using AMSTAR 2) and the limited area of research topics, future high-quality original studies and meta-analyses should be conducted to evaluate the associations between nutritional, lifestyle, and metabolic factors and NAFLD.The raw data supporting the conclusions of this article will be made available by the authors, without undue reservation.YZ, KN, and YX contributed to the study conception and design. YX, QW, HD, JL, YL, HS, YJ, QC, and KN contributed to check the data and results. YX and QW contributed to the drafting and revising of the manuscript. KN and YZ contributed to the approval of the final version of the manuscript. All authors contributed to the article and approved the submitted version.This study was supported by grants from the National Key R&D Program of China (No. 2017YFC0907401), the National Natural Science Foundation of China (No. 81903302 and No. 91746205), the LiaoNing Revitalization Talents Program (No. XLYC1907102 and No. XLYC1802095), China Postdoctoral Science Foundation (No. 2018M641752 and No. 2018M641753), Shenyang high level innovative talents support program (No. RC190484), and 345 Talent Project of Shengjing Hospital of China Medical University (No. M0294 and No. M0268).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.


Acute radiation dermatitis (ARD) is the most common acute response after adjuvant radiotherapy in breast cancer patients and negatively affects patients’ quality of life. Some studies have reported several risk factors that can predict breast cancer patients who are at a high risk of ARD. This study aimed to identify patient- and treatment-related risk factors associated with ARD.PubMed, Embase, Cochrane Library, China National Knowledge Infrastructure, and WanFang literature databases were searched for studies exploring the risk factors in breast cancer patients. The pooled effect sizes, relative risks (RRs), and 95% CIs were calculated using the random-effects model. Potential heterogeneity and sensitivity analyses by study design, ARD evaluation scale, and regions were also performed.2 , large breast volume , smoking habits , and diabetes . Of the seven treatment-related risk factors, we found that hypofractionated radiotherapy reduced the risk of ARD in patients with breast cancer compared with that in conventional fractionated radiotherapy . Sequential boost and bolus use was significantly associated with ARD . However, chemotherapy regimen , hormone therapy , trastuzumab therapy , and nodal irradiation  were not correlated with ARD. Sensitivity analysis results showed that BMI was consistently associated with ARD, while smoking, breast volume, and boost administration were associated with ARD depending on study design, country of study, and toxicity evaluation scale used. Hypofractionation was consistently shown as protective. The differences between study design, toxicity evaluation scale, and regions might explain a little of the sources of heterogeneity.A total of 38 studies composed of 15,623 breast cancer patients were included in the analysis. Of the seven available patient-related risk factors, four factors were significantly associated with ARD: body mass index (BMI) ≥25 kg/m2 was a significant predictor of ARD and that hypofractionation was consistently protective. Depending on country of study, study design, and toxicity scale used, breast volume, smoking habit, diabetes, and sequential boost and bolus use were also predictive of ARD.The results of this systematic review and meta-analysis indicated that BMI ≥ 25 kg/m Breast cancer is the most common malignancy in women , 2. Due RT targets tumor cells and induces double-stranded DNA breaks, resulting in cell damage and death, as well as damage to the surrounding normal tissue . Due to Consequently, we believe that a high-quality systematic review and meta-analysis is needed to summarize currently available data to obtain an exact conclusion. This systematic review and meta-analysis aimed to evaluate the risk factors that are significantly associated with acute ARD in women with breast cancer and provide more evidence for the prevention and management of ARD.https://www.crd.york.ac.uk/prospero/display_record.php?ID=CRD42021250289)  guidelines. The study protocol was registered in PROSPERO  and two Chinese databases  were searched from January 2000 to May 2021. A manual search of the reference lists of the identified literature and systematic reviews was also conducted. Only articles published in English or Chinese were included. Based on a combination of MeSH terms and keywords, the following research terms were used: “breast cancer,” “radiotherapy,” and “radiation dermatitis.” The search strategy is shown in The included study needed to meet the following criteria: 1) the research participants were breast cancer patients aged 18 years and older undergoing radiotherapy; 2) the purpose of the study was to assess patients’ skin toxicity reactions and tumor- and treatment-related factors that increase the risk of radiation-induced acute skin toxicity in breast cancer patients; 3) the study outcomes were the prevalence, incidence, and severity of acute skin reactions (radiation dermatitis and erythema) induced by radiotherapy; 4) the study design was a randomized controlled trial (RCT) or observational study design, including cohort and case–control studies; and 5) relative risks (RRs), odds ratios (ORs), and hazard ratios (HRs) were used as measures of effect.Studies were excluded if they were books, reviews, case reports, experimental laboratory articles, conference abstracts, opinion articles, commentaries, and editorial reviews.Study selection and data extraction were performed by two independent authors using the PRISMA flow diagram. Dissent was resolved by discussion or consultation with a third author. When data were incomplete, the original authors were contacted.p-values were also gathered.The following data were extracted for each article: first author, country, publication year, study design, study period, patient characteristics, follow-up duration, sample size of participants, all risk factors investigated, and outcome measured. Finally, the adjusted OR, RR, and HR, and 95% CIs and The Cochrane Handbook for Systematic Reviews was used to assess the quality of RCTs. Quality scoring of the observational study was performed using the Newcastle–Ottawa Scale. Funnel plots and Egger’s tests were performed to assess publication bias, in which Egger’s regression test  was perfI2 statistics were conducted to assess the potential heterogeneity between individual studies, with values of the latter above 75% being regarded as indicating high levels of heterogeneity. A sensitivity analysis was also carried out using the “leave-one-out” method.To determine the risk factors associated with ARD, the Stata version 16 software was used for data synthesis. The RR and 95% CI of the outcome variables were calculated. The pooled effect sizes of the studies were visualized using a forest plot. The random-effects model was applied to calculate the pooled RR and its 95% CI if significant heterogeneity among studies was found. Otherwise, a fixed-effects model was used. Sensitivity analysis of the study design (RCT and cohort), ARD assessment scale (Radiation Therapy Oncology Group (RTOG) and Common Terminology Criteria for Adverse Events (CTCAE)), and area  was performed. Cochran’s Q test and A total of 793 articles were initially identified through database searches after removing 105 duplicate articles. The abstracts and titles of 606 articles were reviewed, and 573 articles were excluded. The remaining 123 articles were read and screened in full texts for further assessment of eligibility. A total of 85 articles were further excluded for the following reasons: review or meta-analysis, small sample study, risk factors that are rarely studied, conference abstracts, non-intended endings, study being repeated on the same population, and lack of available data. Finally, 38 studies were included in this meta-analysis, of which five studies were RCTs, nine studies were retrospective designs, and 24 studies were prospective design , 23–53. The characteristics of the included studies are summarized in 2 , large breast volume , smoking habits , or diabetes  had significantly higher risks of ARD than their counterparts. However, a significantly increased risk was not observed with increasing age , hypertension , or race .Random-effects meta-analyses were conducted for patient-related risk factors, including age, body mass index (BMI), breast volume, smoking, race, hypertension, and diabetes, as shown in I2 statistics, a substantially large inconsistency  was found for significant heterogeneity among studies regarding the breast volume. BMI  and smoking habits  showed moderate inconsistency with significant heterogeneity in each meta-analysis. No heterogeneity was found for hypertension, diabetes, or race. There was no indication of publication bias, as implied by the funnel plot and Egger’s tests for the risk factors of age (p = 0.084) and smoking habits (p = 0.284). However, funnel plots and Egger’s test indicated potential publication bias for BMI (p = 0.016) and breast volume (p = 0.001) , sequential boost , hormone therapy , HFRT versus CFRT , trastuzumab therapy , and nodal irradiation . No statistically significant heterogeneity was detected for bolus . No evidence of asymmetry in the funnel plot was found, and Egger’s tests also showed no significant evidence of publication bias for chemotherapy (p = 0.676), hormone therapy (p = 0.152), or HFRT versus CFRT (p = 0.07). However, a potential publication bias was observed for boosts in the funnel plot and Egger’s test (p = 0.002) , were related to ARD. Age, hypertension, chemotherapy, hormone therapy, trastuzumab therapy, and nodal irradiation were not associated with radiation dermatitis.p < 0.0001) (The 10-year follow-up of the UK Standardisation of Breast Radiotherapy (START) trials confirmed that appropriately dosed HFRT was safe and effective in patients with early breast cancer –57. Norm 0.0001) . A meta- 0.0001) . Another 0.0001) . A large 0.0001) . Consist 0.0001) , 61. HFR 0.0001) .To improve the local control rate, it is necessary to administer a localized dose escalation (boost) to the tumor bed. However, as our results show, sequential application of a boost increases the risk of ARD. Therefore, it is urgent to find a suitable boost administration method that reduces the risk of side effects without compromising local control. A study found that patients with simultaneous integrated boost had lower toxicity than those receiving a sequential boost or no boost . Two revOur results showed that nodal irradiation was not associated with ARD, but nodal irradiation resulted in a larger irradiated volume. Regarding the dose distribution of the target volume and skin and the occurrence of skin toxicity, existing data are sparse. One prospective study found that subclavian skin volume is correlated with medium-term skin toxicity . Two othPatient-related risk factors, such as BMI, breast volume, smoking habits, and diabetes, were found to be risk factors for ARD. Large breast volume and high BMI have been most frequently reported to increase the risk of ARD. BMI is strongly related to breast volume . Breast It was necessary to investigate potential sources of heterogeneity. First, the type of research design included in the study is different, resulting in substantial heterogeneity. Therefore, a sensitivity analysis was conducted according to the type of study . For the two risk factors of breast volume and boost, the results of the prospective cohort study and RCT were conflicting, proving the heterogeneous results of the study design. Second, acute toxicity was evaluated using the most common tools: the NCI CTCAE or RTOG scale. One study found a high concordance between the RTOG and CTCAE criteria (correlation coefficients >0.9) . NeverthIn addition to the large heterogeneity of our results, there are other shortcomings that need to be considered. First, the number of included studies was limited, especially for some risk factors ; therefore, it is insufficient for statistical analyses, and the results should be interpreted with caution. Second, cohort studies are the main part of the included studies and carry inevitable inherent biases. Cohort studies do not use randomization; therefore, the groups may not be comparable, leading to selection bias. In addition, not all studies adjusted for confounding factors and recall bias due to selective reports, or the presentation of incomplete result data may affect the results of the analysis. These are potential deviations that may affect the validity of the research results. Third, this article only studied the risk factors of ARD; other side effects of radiotherapy, such as late radiation dermatitis, radiation pneumonitis, and radiation esophagitis, still need to be further explored. Finally, studies have reported that genetic analysis can predict patients who are at a higher risk of ARD , 76. How2 was a significant predictor of ARD, and hypofractionation was consistently protective. Depending on country of study, study design, and toxicity scale used, breast volume, smoking habit, diabetes, and sequential boost and bolus use were also predictive of ARD. On the basis of this study, doctors could predict patients with breast cancer at high risk of ARD at the outset of treatment options, adjust treatment plans, and take necessary precautions. In the future, more accurate predictions, such as genetic markers, are expected.In conclusion, BMI ≥ 25 kg/mThe raw data supporting the conclusions of this article will be made available by the authors, without undue reservation.The authors’ responsibilities were as follows: YX and JC contributed to the conception and design of the research. YX and QW extracted the data and performed the analyses. YX, QW, and TH interpreted the evidence and wrote the manuscript. RC, JW, and HC revised the article. All authors contributed to the article and approved the submitted version.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.


Objective: To evaluate the risk factors of perioperative hemodynamic instability in pheochromocytoma, we conducted a systematic search of the literature using the Preferred Reporting Items for Systematic Reviews and Meta-analysis. Methods: In April 2021, we systematically searched PubMed, the Cochrane library, and Scopus for relevant studies on the risk factors of perioperative hemodynamic instability of adrenalectomy in patients with pheochromocytoma, and we subjected the findings from those studies to formal meta-analysis. Results: Our systematic review identified 14 studies involving 1725 patients, of which nine studies with 967 patients were eligible for meta-analysis. The results of meta-analysis showed that tumor size (odds ratio (OR): 1.14 for each increased cm, 95% confidence interval (CI) 1.03–1.26, z = 2.57) and urinary norepinephrine  were most closely associated with the occurrence of perioperative hemodynamic instability. Conclusion: These findings suggest that tumor size and urinary norepinephrine are important predictors and risk factors for perioperative hemodynamic instability in adrenalectomy for pheochromocytoma. Such findings may be of value to surgeons and anesthesiologists when considering or preparing for this procedure. Adrenal pheochromocytoma is defined as a tumor derived from catecholamine-producing chromaffin cells in the adrenal medulla. Almost all adrenal pheochromocytomas produce, store, release, and metabolize catecholamines and can cause life-threatening systemic effects such as, stroke, heart attack, and multiple organ failure .Although resection is the only curative treatment for pheochromocytoma, this surgery carries a very high risk of eliciting massive catecholamine release, which can cause severe hypertension ,3. HypotTo date, although several researchers have investigated possible risk factors for perioperative hemodynamic instability in pheochromocytoma, no systematical evaluation has been performed ,6,7,8. WOur systematic review and meta-analysis were based on the requirements of the Preferred Reporting items for Systematic Review and Meta-analysis (PRISMA) statement . The proEligible studies were those that compared pheochromocytoma patients who experienced hemodynamic instability to those who did not, either during or after surgery, with the objective of assessing the relationship between risk factors and hemodynamic instability, utilizing univariate and multivariate logistic regression analysis in cohort studies. We excluded articles that were published in a language other than English, reviews, commentaries, and case series. If multiple articles were published by the same group using similar cohorts, we selected either the more recent or the higher quality publication.Two investigators (F.U. and S.K.) worked independently to extract the required data. Data included the first author’s name, publication year, country in which patients were enrolled, period of enrollment, number of patients, age, tumor size, and risk factors. Odds ratios (ORs) and 95% confidence intervals (CIs) were determined for risk factors associated with occurrence of hemodynamic instability. All discrepancies related to data extraction were resolved in a consensus meeting.After selecting the studies for inclusion, we assessed the quality of each study on the Newcastle-Ottawa Scale , based op-value, we calculated the 95% CI [2 statistics were used to evaluate heterogeneity among outcomes of the studies in this meta-analysis, with significant heterogeneity indicated by p < 0.05 in the Cochrane Q test and ratio >50% in I2 statistics and with the use of random effect models based on the DerSimonian and Laird method [p < 0.05.A forest plot was used to assess ORs from the multivariate logistic regression analyses of individual studies and to obtain a summary OR for the relationship between risk factors and hemodynamic instability. If the study reported only the OR and e 95% CI ,13. The d method ,15,16. WWe identified a total of 923 articles from the search query. Of those, 72 duplicates were removed, and 770 articles were excluded after initial screening and abstract review. Sixty-seven additional articles were excluded after full-text evaluation. The remaining 14 articles were subjected to systematic review ,25,26, aWe conducted a meta-analysis of baseline patient and tumor characteristics factors such as patient age, body mass index (BMI), tumor size, and urinary norepinephrine. Tumor size, age, and BMI were evaluated as continuous variables in the meta-analysis. Urinary norepinephrine was evaluated as a categorical variable.2 = 31.36; p < 0.001) and I2 test (I2 = 84.1%) showed significant heterogeneity. The funnel plot identified one study over the pseuo-95% CI (Six studies (693 patients) provided data on the relationship between tumor size and hemodynamic instability. The forest plot a showed o-95% CI a.2 = 2.75; p = 0.098) and I2 test (I2 = 63.6%) revealed no heterogeneity, so we used a fixed-effect model. The forest plot (Two studies (119 patients) provided data on the relationship between urinary norepinephrine and hemodynamic instability. The Cochrane Q test  and I2 test (I2 = 4.0%) revealed no heterogeneity, so we used a fixed-effect model. The forest plot (Two studies (257 patients) provided data on the relationship between age and hemodynamic instability. The Cochrane Q test  and I2 test (I2 = 88.1%) revealed significant heterogeneity. The funnel plot identified one study over the pseuo-95% CI (Two studies (386 patients) provided data on the relationship between BMI and hemodynamic instability. The forest plot d showed o-95% CI d.Surgical procedure (open surgery) , retropeTo the best of our knowledge, our study is the first systematic review and meta-analysis to investigate the risk factors for perioperative hemodynamic instability in pheochromocytoma. Our meta-analysis utilized data from nine published articles with a combined patient population exceeding 900 patients. We used those data to evaluate factors associated with perioperative hemodynamic instability.First, we found that tumor size (continuous variable) was associated with perioperative hemodynamic instability. Adrenal pheochromocytomas produce, store, release, and metabolize catecholamines, and larger pheochromocytomas involve considerable endocrine activity, which can result in severe hypertension during adrenalectomy. Notably, the resection of larger tumors has also been associated with chronically low circulating blood volume and an abrupt decrease in serum levels of catecholamines, which can cause severe hypotension after adrenalectomy. Moreover, in adrenalectomy, a larger tumor size entails difficulty and increases the degree of manipulation necessary to remove the tumor, which can also cause hemodynamic instability during operation . To dateWe also found that urinary norepinephrine was associated with perioperative hemodynamic instability. Additionally, although we could not perform meta-analysis, previous studies have shown that urinary epinephrine can also be a risk factor for hemodynamic instability ,19. AdreWe found several additional factors, including preoperative beta-blockade therapy and the surgical approach, were significantly associated with perioperative hemodynamic instability. However, those factors were excluded from meta-analysis because they were evaluated in only one study. In pheochromocytoma, an alpha blocker is generally used prior to pheochromocytoma resection, but the initiation of beta-blocker administration is sometimes considered for additional blood pressure control and control of tachyarrhythmias . ThompsoLaparoscopic adrenalectomy has been shown to be a safe and feasible procedure for pheochromocytoma  and can This study represents the first systematic review and meta-analysis to assess risk factors for hemodynamic instability in patients with pheochromocytoma. The study has several limitations, however. First, selection (reporting) bias might lead to less frequent publication of negative findings. All studies in our meta-analysis were retrospective in design, increasing the risk of selection bias. Second, there was no consensus on cut-off values for the risk factors selected in our study. Most investigators selected their cut-off values based on their preferred statistical methods or on independently pre-defined biomarker cut-off values from the literature. Third, the studies did not use a uniform definition of perioperative hemodynamic instability. Indeed, vasoactive agents and volume therapy can directly influence the definition but usually not considered. In this context, recently, the hemodynamic instability score was proposed to quantify the overall degree of hemodynamic instability, and it may have future applications in both patient management and clinical research . Fourth,Our meta-analysis indicates that tumor size and urinary norepinephrine are closely related to the occurrence of hemodynamic instability in patients with pheochromocytoma. These findings may be helpful to surgeons and anesthesiologists in cautiously preparing for perioperative hemodynamic instability in these patients.


The efficacy and safety of impregnated central venous catheters (CVCs) in pediatrics remain controversial. The purpose of this study was to evaluate the efficacy of impregnations for the prevention of catheter-related bloodstream infection (CRBSI).I2 statistics, and sensitivity analysis and subgroup analysis were also performed.We searched the following five electronic databases: Medline, PubMed, Cochrane, Embase, and the Web of Science for randomized controlled trials (RCTs) up to March 2021. Pooled risk ratios (RRs) with 95% confidence intervals (CIs) were calculated using a fixed-effects model. Assessment of publication biases was evaluated by Egger's test. Heterogeneity between studies was assessed based on the chi-square test and A total of six RCTs with 3,091 patients were included. Impregnated CVCs provided significant benefits in reducing the risk of CRBSI  in pediatric patients, especially in the pediatric group. No publication bias was observed in the Egger test for the risk of CRBSI. Drug type is a source of heterogeneity.Antimicrobial-impregnated CVCs are beneficial to prevent CVC-related complications in pediatrics. Central venous catheters (CVCs) are clinically important in the treatment of venous pressure monitoring, infusion of drugs, and nutrient fluid supplementation . HoweverThe prevention strategies of CRBSI being recommended include highlighting hand hygiene , maximalGiven this background, we performed a meta-analysis study of available studies to evaluate the efficacy of central venous catheters impregnated with heparin and antimicrobial agents in comparison with standard and non-impregnated catheters on the prevention of CRBSI.We conducted the meta-analysis according to the Preferred Reporting Items for Systematic Review and Meta-Analysis Protocol (PRISMAP) .ClinicalTrials.gov and the WHO International Clinical Trials Registry Platform for unpublished, planned, or ongoing trial reports. In addition, we contacted the authors of the included RCTs to obtain additional data if necessary. All studies were imported into EndNote for duplication exclusion.The search was performed up to March 2021 in PubMed, Medline, Embase, the Cochrane Central Register of Controlled Trials, and the Web of Science using MeSH terms and text words for published clinical trials, and the references of relevant meta-analyses and other internet sources for unpublished clinical trials were also considered. The following terms were adjusted according to different search rules in each database: “impregnated,” “catheters,” “pediatrics,” “bloodstream,” and “infections.” We also searched All potential studies were screened based on the PICOS  standard. We defined the terms as follows: (1) Population: all patients with vascular catheters inserted for an expected duration of more than 72 h under the age of 18; (2) Intervention: the intervention method was impregnation-treated CVCs with no limitation of drug types, and the control group refers to conventional, non-impregnated CVCs; (3) Comparison: the objective was to compare the efficacy of impregnated CVCs and non-impregnated CVCs in the prevention of CRBSI in children; (4) Outcomes: the outcomes included the incidence of CRBSIs and thrombosis; (5) Study design: all study designs were prospective RCTs.The primary outcome in this meta-analysis was the incidence of CRBSIs. Furthermore, we also considered the incidence of thrombosis as the secondary outcome with incomplete data among the included studies.Two reviewers (Zhengrong Deng and Jiangwei Qin) independently evaluated the titles, abstracts, and full texts of identified studies using predefined criteria and extracted study characteristics and outcome data for each study, such as first author name, publication year, country, age of population, numbers of participants, interventions, clinical signs, and brief conclusion, with all the data cross-checked by other authors. Any disagreements were resolved to a consensus by team discussion.The risk of bias of the included RCTs was evaluated using the Cochrane Risk of Bias Scale independently using the following criteria: (1) random sequence generation (selection bias), (2) allocation concealment (selection bias), (3) blinding of participants and personnel (performance bias), (4) incomplete outcome data (attention bias), (5) selective reporting (reporting bias), and (6) other sources of bias. For each criterion, studies were classified as having a low, high, or unclear risk of bias.I2 <50%, p > 0.05), a fixed-effects model was used; otherwise, a random-effects model was considered. We set p <0.05 as statistically significant for hypothesis testing. Heterogeneity between studies was also assessed based on the chi-square test by I2 statistics. Assessment of publication biases was evaluated by Egger regression. Sensitivity analysis was conducted to identify the influence of each study on the synthesized results. p <0.05 was considered statistically significant. Subgroup analyses based on impregnation drug type and between neonates (age ≤ 28 days) and children (28 days < age ≤ 18 years) were undertaken to determine the source of heterogeneity. Furthermore, meta-regression analyses were used to detect whether other variables, such as publication year and bloodstream infection (BSI) reported, caused heterogeneity to some extent.The analyses were performed in Review Manager (Version 5.3) and R (Version 4.04). The incidence of CRBSI and thrombosis are binary outcomes, so risk ratios (RRs) with 95% confidence intervals (95% CIs) were calculated using the Mantel–Haenszel method. If there was no statistically significant heterogeneity among the studies . A fixed effects model was used for the meta-analysis because of moderate heterogeneity   . We alsot = −1.05, p = 0.3520 > 0.05).We performed Egger's test of funnel plot asymmetry on the risk of CRBSI in Considering the moderate heterogeneity in the synthetic result, we conducted a sensitivity analysis by excluding 1 study at a time to estimate whether the results could have been affected markedly by a single study . The resI2 = 26%, p = 0.25), while impregnated heparin showed higher heterogeneity , indicating that the efficiency of heparin impregnated on CRBSI was unstable. In addition, it showed that the antimicrobial-impregnated catheter provided more benefits  in reducing the risk of CRBSI among the included RCTs compared to the heparin-impregnated CVCs.To test our assumption, we conducted subgroup analysis stratified by heparin-impregnated CVCs and antimicrobial-impregnated catheters. As I2 = 51%, p = 0.11 in the neonate group; I2 = 66%, p = 0.09 in the children group). It should be noted that although results with heterogeneity should be considered conservatively, impregnated CVCs seemed to protect children from CRBSI with an RR = 0.37 .We also added an exploratory subgroup analysis regarding the differences in infection rates between neonates (age ≤ 28 days) and children (28 days < age ≤ 18 years) for discrepant immune system development. The results shown in p = 0.253). A fixed model was used because no heterogeneity was noted among the studies . The results are presented in A meta-analysis of three studies revealed no significant differences in the rate of thrombosis between impregnated and standard CVCs , 19, 20 We exhaustively collected RCTs related to impregnated CVCs in pediatrics, which included agents such as heparin, AgION, minocycline, miconazole, and rifampicin. Heparin is a natural anticoagulant substance in animals, and its anti-infective function was suspected to be related to preventing bacterial colonization through beneficial hemodynamic function. However, as we described in the subgroup analysis in It has been reported that pediatric intensive care units (PICUs) have one of the highest rates of acquired bloodstream infections in hospitals, especially after cardiac surgery , with ceTo date, much new evidence has basically unified the perspective that impregnated CVCs are recommended for adults to reduce bloodstream infections. Chong et al. conducted a network meta-analysis including 60 studies with 17,255 catheters and investigated the effects of 14 impregnation drugs on CRBSI and catheter colonization. Significant CRBSI reduction was associated with minocycline-rifampicin (RR = 0.29 [95% CI: 0.16–0.52]) and silver (RR = 0.57 [95% CI: 0.38–0.86]) impregnations compared to no impregnation. For colonization, miconazole-rifampicin (RR = 0.14 [95% CI: 0.05–0.36]), 5-fluorouracil (RR = 0.34 [95% CI: 0.14–0.82]), and chlorhexidine-silver sulfadiazine (RR = 0.60 [95% CI: 0.50–0.72]) impregnations showed significant decreases compared with non-impregnation . Wei et p = 0.0002). Our further subgroup analysis indicated that the use of antimicrobial drug-impregnated CVCs had a significantly lower risk of CRBSI (RR = 0.48) than the conventional catheter group in the pediatric population. In addition, after considering differences between neonates and children, the age cutoff at 28 days supported a prominent prevention effect in the pediatric group (28 days < age ≤ 18 years). Our results support the benefits antimicrobial drug-impregnated CVCs would have in pediatric patients, especially in children.Our systematic review and meta-analysis provided the most recent and comprehensive analysis of the efficiency of impregnated CVCs for the prevention of CRBSI in pediatrics. With six RCTs included, the results of this meta-analysis suggest that the use of impregnated CVCs is beneficial to reduce the risk of CRBSI for patients with CVC and is an effective anti-infection strategy in preventing CRBSI .The datasets generated for this study are available on request to the corresponding author.YM, ZD, and JQ: contributed to the study conception, design, data acquisition, and interpretation.YM, ZD, JQ, HS, and FX: contributed to the data analysis and drafted the manuscript. All authors contributed to the critical revision of the manuscript and approved the final version.This study was supported by a grant from the West China Second University Hospital of Sichuan University, KX144, which funded the design of the study and collection, analysis, and interpretation of data and the writing of the manuscript.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.


Does neuraxial labor analgesia (NLA) increase the risk of autism spectrum disorder (ASD) in the offspring?Using a birth cohort of more than 1.6 million children nested in nationwide health care utilization data with as long as 10 years of follow-up data, evidence of a strong association between NLA and ASD was not found after accounting for potential confounders. Meta-analysis of all recently published data yielded similar results.While a small increase in risk cannot be excluded, given the possibility of some residual confounding, these findings do not support the notion that NLA is associated with an increased risk of ASD. This cohort study evaluates whether neuraxial labor analgesia is associated with increased risk of autism spectrum disorder in the offspring. Recent studies have reported conflicting findings regarding a potential association between analgesia used during labor and autism spectrum disorder in the offspring.To evaluate whether neuraxial labor analgesia increases the risk of autism spectrum disorder in the offspring.This cohort study included mother-child dyads who underwent vaginal delivery and were exposed to neuraxial labor analgesia. Delivery data were collected from the Medicaid Analytic eXtract (2005-2014) for mothers with public insurance and the IBM Health MarketScan Research Database (2005-2015) for mothers with private insurance. Data analysis was conducted from January to October 2021.Presence of a procedure code indicating neuraxial labor analgesia.Children with autism spectrum disorder, identified using a validated algorithm . Cumulative incidence curves stratified by exposure were assessed using Kaplan-Meier analyses. Hazard ratios were estimated through Cox proportional hazards regression, using propensity-score fine stratification for confounding control. Estimates from both insurance cohorts were combined through fixed-effects meta-analysis. Subsequently, results from these analyses were combined with existing published studies.The cohort of mother-child dyads with public insurance consisted of 910 696 deliveries , with 484 752 (53.2%) being exposed to neuraxial labor analgesia. The cohort of mother-child dyads with private insurance included 696 883 deliveries , with 513 347 (73.7%) being exposed. Cumulative incidence of autism spectrum disorder by 10 years of age was 1.93%  among children in the exposed group vs 1.64%  among children in the unexposed group in the publicly insured cohort. Respective numbers were 1.33%  and 1.19%  in the privately insured cohort. Adjusting for potential confounders and pooling across both cohorts resulted in a hazard ratio of 1.08 . Results were consistent when additionally adjusting for empirically identified variables through high-dimensional propensity score analyses  or expanding the cohorts to include cesarean deliveries and assisted vaginal deliveries . Meta-analysis of this study and recently published observational studies yielded similar findings with a pooled hazard ratio of 1.10 .Although a small increase in risk cannot be ruled out, the combined evidence from observational studies does not support the notion that neuraxial labor analgesia is associated with an increased risk of autism spectrum disorder. While its etiology remains unknown, there is some evidence suggesting that—apart from its inheritable component5 the finding of an increased risk of ASD garnered substantial attention. Five US medical societies (together representing more than 100 000 physicians) as well as the British Royal College of Anaesthetists expressed strong concern that risk estimates might be biased, as important information on pregnancy and delivery complications, which are known to increase the risk of ASD and are also associated with an increased rate of epidurals, was not accounted for.7Given that epidural labor analgesia—and more generally, neuraxial labor analgesia (NLA)—is the widely accepted standard for effectively providing pain relief throughout labor, with most women in the United States using this method,10 All reported risk estimates that were much lower than the previously observed 37% increase in risk. As these conflicting results may cause confusion and unease in pregnant women and their health care practitioners regarding the use of labor epidurals, it is crucial to obtain and communicate as much reliable information as possible about the safety of NLA and to summarize all available evidence.Subsequently, 3 additional studies have examined this association.Using information from 2 large population-based health care utilization databases that together contain information on women with both public and private insurance in the United States and that have rich information on potential confounders, along with a validated algorithm to identify ASD, we evaluated the association between NLA exposure and the risk of ASD in the offspring. Furthermore, we conducted a meta-analysis of all published studies focusing on this association to evaluate the totality of scientific evidence.11 As Medicaid currently covers approximately 50% of births throughout the United States12 and commercial insurers cover most of the balance, these 2 cohorts together are expected to be highly representative of the nationwide obstetric population. Mothers were required to be insured from 3 months before pregnancy until 1 month after delivery to ensure complete covariate ascertainment. Start and length of pregnancy was estimated using a previously validated algorithm based on diagnostic codes for preterm birth.13 Children were followed up until ASD diagnosis, end of insurance enrollment, or end of the study period, whichever came first. Cesarean deliveries and assisted vaginal deliveries  were excluded from the main analyses but were included in sensitivity analyses . The exclusion and inclusion criteria are described in We conducted a cohort study of mothers with public and private insurance linked to their liveborn children nested in the Medicaid Analytic eXtract (MAX) from 2005 to 2014 (the most recent years available at the time of study conduct) and the IBM Health MarketScan Research Database (MarketScan) from 2005 to 2015. The MAX database consists of health care claims from the nationwide Medicaid  populations and presents a racially diverse and relatively young population. The MarketScan database is the largest nationwide data set in the United States that is based on commercial health insurance claims. Both data sources provide rich patient-level information on demographic characteristics, insurance enrollment, outpatient medication dispensing, outpatient and emergency department visits, and hospitalizations as well as their accompanying diagnoses and procedures. For both cohorts, we have previously developed an algorithm to link mothers to their liveborn children based on state, insurance case number (to identify family units), and date of delivery.STROBE) reporting guideline for observational studies.The use of the deidentified databases for research was approved by the institutional review board at Brigham and Women’s Hospital, which waived the need for informed consent. This report follows the Strengthening the Reporting of Observational Studies in Epidemiology , procedure codes 03.90 and 03.91 or Current Procedural Terminology (CPT) codes 01967, 62319, and 00955. The reference group included all women who neither had a code indicating NLA nor a code indicating other types of anesthesia used close to or during delivery .Exposure was based on the presence of any of the following codes indicating NLA: ICD-9 codes 299.x, pervasive developmental disorders, excluding 299.1x, childhood disintegrative disorder) at 1 year or older. This algorithm has been shown to identify the outcome of ASD with a high positive predictive value based on medical record review .14Children with ASD were identified based on the presence of at least 2 medical encounters with a documented diagnosis of ASD , obstetrical and labor complications , maternal comorbidities , prenatal medication exposure , and county-level information on socioeconomic status measures  (available in MAX only). For a detailed description on the individual variables and assessment periods, please refer to the eTable in the 15Analyses were conducted separately for the publicly and privately insured cohorts. Hazard ratios (HRs) were combined using fixed-effects meta-analysis.X̄ represents the sample mean and s2 the sample variance of the covariate in the exposed (exp) and the reference group (ref)  as well as mediators on the causal path from NLA to ASD. If they are mediators, adjustment would result in an estimate of the direct effect of NLA on ASD rather than the total effect. Thus, to account for both scenarios, analyses were conducted with and without adjustment for these conditions.Unadjusted and adjusted HRs and 95% CIs were estimated using Cox proportional hazards regression. Adjustment for potential confounders or proxies for confounders was done using propensity score (PS) fine stratificationIn confirmatory analyses, to account for potential residual confounding, we used high-dimensional PS analyses to identify 200 empirically defined covariates in addition to the prespecified variables included in level 2. Here again, 2 models were created: model 3 included empirical variables assessed from 3 months prior to pregnancy until the day before delivery, and model 4 included variables assessed from 3 months before pregnancy to 1 month after delivery to increase the capture of potential risk factors for ASD occurring or being recorded around the time of or shortly after delivery, similar to the variables included in model 2.19 Finally, the study by Qiu and colleagues3 that reported on an increase in risk associated with NLA also did not include cesarean deliveries, and our goal was to assess whether we could replicate their findings. Nevertheless, because cesarean deliveries represent such a large proportion of deliveries in the United States and are generally done under regional anesthesia, we conducted a sensitivity analysis including cesarean deliveries and other operative deliveries.We excluded cesarean deliveries from the main analyses for several reasons. First, most cesarean deliveries in the United States are performed using neuraxial anesthesia, raising issues around defining a relevant comparator group . Second, the brief duration of exposure to local anesthetics in the context of cesarean delivery would not be expected to confer the same risk as what has been posited for labor epidurals. Third, cesarean deliveries might directly or indirectly affect the neurological development in the offspring through other pathways, such as altered microbiota or prenatal complications that result in these interventions.epidural[Title]) OR  OR  OR (anesthesia[Title])) AND (Autism[Title]). Articles were included if they reported HRs for epidural/NLA exposure and ASD. Animal studies, basic science research, case reports, case series, editorials, letters to the editor, and commentaries were excluded. We performed a fixed-effects meta-analysis, using the highest-level adjusted estimates reported in each of the studies, quantified between-study variability using the χ2 test of heterogeneity, and summarized results in a forest plot.Lastly, we conducted a literature review and meta-analysis for published studies focusing on the association of epidural/neuraxial labor analgesia with risk of ASD. We searched MEDLINE via PubMed for terms . Statistical significance was set at the 5% level, and all tests were 2-tailed.The publicly insured cohort consisted of 910 696 pregnancies , of which 484 752 (53.2%) were exposed to NLA. The privately insured cohort included 696 883 pregnancies , with 513 347 (73.7%) being exposed. When comparing those with vs without NLA exposure within cohorts, the strongest imbalances observed in the publicly insured cohort were that women with NLA exposure were more likely to have delivered in the more recent years; to be younger; to be White; to be from the Midwest or South; to have active smoking; to be diagnosed with hyperemesis/vomiting in pregnancy; to be exposed to prescription opioids, antidepressants, or suspected teratogens; to have more outpatient visits; and to have obstructed/long labor . No subsThe mean age at first ASD diagnosis  of the 3548 ASD cases in the publicly insured cohort was 5.2  years among those exposed to NLA (1794 cases) and 4.8  years among those not exposed to NLA (1754 cases) . RespectIncluding cesarean deliveries and assisted vaginal deliveries substantially increased the cohort size to 1 365 461 in the publicly insured cohort and 1 199 813 in the privately insured cohort. Adjusted results were very similar to those from the main analyses  .10 All 4 studies were included in the meta-analysis and were pooled with the results from our analyses. Results from the meta-analysis yielded a pooled adjusted HR of 1.10  ; however, the P value (.82) suggested very low heterogeneity.The PubMed search using our defined search criteria initially yielded 37 citations, of which 15 were directly related to this research question, and 4 presented original data.06-1.13)  with a PUsing 2 nationwide mother-child linked birth cohorts that together included more than 1.6 million children born vaginally, we did not find evidence of a strong association between NLA exposure and ASD risk. Estimates were consistent when expanding the cohort to include children born via cesarean deliveries or assisted vaginal deliveries. Results were also consistent when meta-analyzing all data published so far, with the upper bound of the 95% CI excluding a more than 13% increase in risk. Therefore, the combined evidence from these observational studies does not indicate a strong association between NLA and ASD.3) to evaluate such an association reported a 37% increase in risk with epidural analgesia during labor. However, the 3 cohort studies that were published shortly thereafter all reported adjusted risk estimates similar to those observed in our data, with HRs ranging from 1.05 to 1.09 and the highest upper confidence limit at 1.20.10 While the estimate from the meta-analysis shifted only slightly toward the null (from 1.10 to 1.07) when excluding the study by Qiu and colleagues,3 statistical testing no longer pointed toward heterogeneity between studies, confirming that the study by Qiu et al3 is indeed substantially different from the others. The explanation for the increase observed in that first study is not clear; residual confounding particular to that study population  is a possibility. The biological plausibility of an association between epidural labor analgesia and ASD is questionable because the diluted local anesthetics and opioids are infused into the mother’s epidural space at such low doses that a fetal neurotoxic effect appears very unlikely.20The first large observational study  than the biggest cohort study published to date focusing on the association of interest and are expected to be broadly representative of the US obstetric population. The large size coupled with rich information on a broad range of covariates and the use of a validated outcome definition allowed for precise risk estimation with robust confounding adjustment.21 ASD cases that were either not recorded or did not meet our definition were missed, thus resulting in a potential underestimation of the true number of children with ASD. However, the ASD incidences observed in our cohorts align well with autism prevalence figures reported by the US Centers for Disease Control and Prevention.24 Furthermore, we opted for a highly specific outcome definition, which will result in nonbiased or minimally biased HRs assuming nondifferential sensitivity. As information in our databases is based on claims used for billing, health care professionals have a strong incentive to provide complete information and are regularly audited to ensure they do not report diagnoses or procedures that do not exist, which would indicate fraud. Because of this, coding of NLA is expected to be accurate. Moreover, as a prerequisite of this study, we explored the feasibility of using the selected ICD-9 and CPT codes by reviewing maternal claims profiles and comparing our estimated NLA prevalence to existing literature25 and found that these codes seemed to adequately capture NLA use. Residual confounding is always a concern in nonrandomized studies, and the small increase in risk we observed is likely attributable to residual confounding. We were not able to address a potential dose-response association, as our data do not contain information on NLA duration and dosage. Lastly, we restricted our main analyses to vaginal deliveries because our data did not differentiate between neuraxial analgesia prior to or during cesarean delivery nor did they allow us to adequately distinguish different anesthesia methods used during cesarean delivery. Nevertheless, this restriction is not expected to result in selection bias because NLA has been shown not to increase the risk of cesarean deliveries.26 Furthermore, when including cesarean delivery and assisted vaginal deliveries, adjusted results were very consistent with those from our main analyses.This study also has limitations. As both outcome and exposure are based on the presence of specific diagnostic and procedure codes, we need to equate absence of a claim to absence of a condition or treatment. While our algorithm to identify ASD has yielded a high positive predictive value,28 and NLA is the widely accepted standard for effective and safe labor analgesia used by most pregnant women in the United States, these findings are important and reassuring for both pregnant women and their health care practitioners when considering options for labor pain relief.Overall, findings from our study as well as the results from the meta-analysis do not suggest a strong association of NLA use with risk of ASD. While we cannot exclude a small increase in risk, given the observational nature of these types of studies, this small elevation can easily be explained by a small amount of residual confounding related to differences in underlying maternal, obstetric/labor, and genetic factors. Given that childbirth is typically the most painful moment in a woman’s life, maternal pain can lead to severe consequences ,


FGF20 gene that has been implicated in PD. The variation of FGF20 in the 3′ untranslated region was shown to increase α-synuclein expression. We examined the association of rs591323 with the risk of PD in a Taiwanese population and conducted a meta-analysis, including our study and two other studies from China, to further confirm the role of this SNP in Taiwanese/Chinese populations. A total of 586 patients with PD and 586 health controls (HCs) were included in our study. We found that the minor allele (A) and the AA + GA genotype under the dominant model are significantly less frequent in PD than in controls. The meta-analysis consisted of 1950 patients with PD and 2073 healthy controls from three studies. There was significant association between rs591323 and the risk of PD in the additive  and the dominant models . Our study results and the meta-analysis support the possible protective role of the rs591323 A allele in PD in Taiwanese/Chinese populations.Parkinson’s disease (PD) is a neurodegenerative disease with the pathological hallmark of Lewy bodies and Lewy neurites composed of α-synuclein. The SNP rs591323 is one of the risk loci located near the  In adFGF20 gene [FGF20 in the 3′untranslated region was shown to increase the expression of α-synuclein [The SNP rs591323 is located in an intergenic region in 8p22 near the F20 gene , which iF20 gene . Interesynuclein , whereasIn this study, we aimed to examine the association of rs591323 with the risk for PD in a Taiwanese population and conduct a meta-analysis to further confirm the role of this particular SNP in Taiwanese/Chinese population.Patients who fulfilled the UK PD Society Brain Bank Clinical diagnostic criteria  were rechttp://www.pdgene.org/gwas (accessed on 4 April 2021)). We used the Agena MassARRAY platform with iPLEX gold chemistry  for genotyping of the 20 SNPs following the manufacturer’s protocol. We designed the sequence of the polymerase chain reaction (PCR) primers and the extension primers with the Assay Designer software package (v.4.0). In brief, a multiplex PCR reaction was performed in a total volume of 5μL that consisted of 1μL DNA sample (10 ng/μL), 1 unit of Taq polymerase, 500 nmol of each PCR primer mix and 2.5 mM of each deoxynucleotide . Thermocycling was performed in the following sequence: 94 °C for 4 min, 45 cycles of 94 °C for 20 s, 56 °C for 30 s, 72 °C for 1 min, and finally 72 °C for 3 min. In total, 0.3 units of shrimp alkaline phosphatase was used to deactivate unincorporated deoxynucleotides. iPLEX enzyme, terminator mix, and extension primer mix were used for the single base extension reaction with thermocycling ). A cation exchange resin was used to remove residual salt from the reactions. In brief, the clean resin was spread out on the dimple plate using the scraper and the resin was left to dry for 10 min at room temperature. Each iPLEX Gold extension product was added with 16 µL high performance liquid chromatography-grade water. The sample plate was sealed and centrifuged at 3200× g for 1 min. The sample plate was gently inverted on the top of the dimple plate. The sample plate wells were aligned to the well of the resin, and then the both plates were inverted again so that the dimple plate was on the top of the sample plate. The dimple plate was gently tapped to let the resin fall into the sample wells and then the dimple plate was removed. The sample plate was sealed and rotated for at least 15 min at room temperature to remove the salt. The purified primer extension reaction was analyzed on a matrix pad of a SpectroCHIP using a MassARRAY Analyzer 4. The TYPER 4.0 software was used for clustering analysis.The FGF20 SNP rs591323 is among the top 20 PD risk foci identified by GWAS meta-analysis in the PDGene database , the pooled OR was calculated using the fixed effect model (the Mantel–Haenszel method). Z test was used to determine the significance of the pooled OR.We examined the effect of the SNP to the risk of PD in the additive model , dominant model (AA + AG versus GG), recessive model (AA versus AG + GG) using the odds ratio (OR) with 95% confidence intervals (CIs). Heterogeneity between the studies was determined by a Q test and IFGF20 and rs591323 in this study. The mean age of the patients with PD was older than the HCs, but not significantly . Gender distribution is similar between groups (p = 0.06). The distribution of all the genotypes followed the Hardy–Weinberg equilibrium (p = 0.011) and the A allele is less frequent in PD than in control (p = 0.029). In the dominant model, the AA + GA genotype frequency is significantly higher in HCs than in PD . When calculated according to the recessive model, there is no difference between the two groups (p = 0.75)  .n = 55). A further 29 studies were excluded through full-text reading to include only the SNP rs591323 and the Taiwanese/Chinese population. Finally, two studies were included for meta-analysis . A total of 1950 PD patients and 2073 HCs were included in the meta-analysis. There was significant association between rs591323 and the risk of PD in both the additive model  and dominant model . The results suggest a possible protective role of the A allele and GA + AA genotype in the dominant model in PD  and a reduction in PD risk for the A allele (p = 0.029) and dominant model (AA + AG versus GG) (p = 0.003). The results of a further meta-analysis support the protective role of the A allele of rs591323 as the results were significant in the additive and dominant models.The SNP rs591323 was first identified as one of risk loci for PD by the International Parkinson’s Disease Genomics Consortium (IPDGC) in 2011 . The resFGF20 gene. Fibroblast growth factor-20 (FGF20) is one of the neurotrophic factors that was found to be selectively expressed in the SNpc [FGF20 genetic variations including one SNP located in the intron (rs1989754) and two SNPs (rs1721100 and ss20399075) located in the 3′ regulatory region are highly associated with the risk of PD [FGF20 remains to be determined. Pak et al. investigated the associations between 19 PD risk loci and dopamine transporter (DAT) and serotonin transporter (SERT) availability in healthy controls. While none of the SNPs had an effect on DAT availability, the AA genotype of rs591323 was associated with lower SERT availability in both pons and midbrain as compared to the AG and GG genotypes [rs591323 is located in the intergenic region in chromosome 8 near the the SNpc , promotithe SNpc . It has sk of PD . Howeverenotypes , the cliIn conclusion, our study is the first meta-analysis to show that the A allele of rs591323 is associated with reduced risk in the Asian PD population. However, the study results should be interpreted carefully with the following limitations of the study in mind. First of all, the sample size was small and only three case-control studies were included for meta-analysis. Second, the average age of the controls was numerically younger than PDs, and conversion of controls to PD at an older age may be possible. Third, the results were not adjusted according to other factors such as patient’s age, sex, and other environmental factors due to lack of information. Lastly, the study result does not directly prove the causal relationship between this particular SNP variation and PD; other possible mechanisms, including gene–gene interactions or linkage disequilibrium with other risk factor genes, may also be involved.