Rehan Qayyum

Relation Between Serum 25-Hydroxyvitamin D and C-Reactive Protein in Asymptomatic Adults (From the Continuous National Health and Nutrition Examination Survey 2001 to 2006)

The inverse relation between vitamin D supplementation and inflammatory biomarkers among asymptomatic adults is not settled. We hypothesized that the inverse relation is present only at lower levels and disappears at higher serum levels of vitamin D. We examined the relation between 25-hydroxyvitamin D [25(OH)D] and C-reactive protein (CRP) using the continuous National Health and Nutrition Examination Survey data from 2001 to 2006. Linear spline [single knot at median serum levels of 25(OH)D] regression models were used. The median serum 25(OH)D and CRP level was 21 ng/ml (interquartile range 15 to 27) and 0.21 mg/dl (interquartile range 0.08 to 0.5), respectively. On univariate linear regression analysis, CRP decreased [geometric mean CRP change 0.285 mg/dl for each 10-ng/ml change in 25(OH)D, 95% confidence interval [CI] -0.33 to -0.23] as 25(OH)D increased ≤21 ng/ml. However, an increase in 25(OH)D to >21 ng/ml was not associated with any significant decrease [geometric mean CRP change 0.05 mg/dl for each 10-ng/ml change in 25(OH)D, 95% CI -0.11 to 0.005) in CRP. The inverse relation between 25(OH)D below its median and CRP remained significant [geometric mean CRP change 0.11 mg/dl for each 10-ng/ml change in 25(OH)D, 95% CI 0.16 to -0.04] on multivariate linear regression analysis. Additionally, we observed a positive relation between 25(OH)D above its median and CRP [geometric mean CRP change 0.06 mg/dl for each 10-ng/ml change in 25(OH)D, 95% CI 0.02 to 0.11) after adjusting for traditional cardiovascular risk factors. In conclusion, from this cohort of asymptomatic adults, independent of traditional cardiovascular risk factors, we observed a statistically significant inverse relation between 25(OH)D at levels <21 ng/ml and CRP. We found that 25(OH)D at a level ≥21 ng/ml is associated with an increase in serum CRP. It is possible that the role of vitamin D supplementation to reduce inflammation is beneficial only among those with a lower serum 25(OH)D.

Heart Rate Variability Predicts ESRD and CKD-Related Hospitalization

Autonomic imbalance, a feature of both diabetes and hypertension, may contribute to adverse cardiovascular outcomes. In animal models, sympathetic nerve activity contributes to renal damage but the extent to which autonomic dysfunction precedes the development of CKD and ESRD in humans is unknown. We measured resting heart rate and heart rate variability in 13,241 adults (45- to 64-years old) followed for a median of 16 years in the Atherosclerosis Risk in Communities (ARIC) Study. We examined heart rate parameters by quartiles, defining those in the lowest quartile (by time and frequency domain measures separately) as the risk group of interest. We identified 199 cases of incident ESRD and 541 patients with CKD-related hospitalizations; higher resting heart rate and lower heart rate variability associated with both outcomes. The fully adjusted hazard ratios for ESRD were 1.98 (95% confidence interval [CI] 1.45 to 2.70) among those in the highest heart rate quartile and 1.56 (95% CI 1.14 to 2.14) for high-frequency power. Other time and frequency domain measures were similarly and significantly associated with ESRD and CKD-related hospitalizations. These results suggest that autonomic dysfunction may be an important risk factor for ESRD and CKD-related hospitalizations and call for further studies to define the mechanisms that underlie these associations.

Systematic Review: Comparative Effectiveness and Safety of Premixed Insulin Analogues in Type 2 Diabetes

Context The relative effects of premixed insulin analogues, other insulin regimens, and noninsulin antidiabetic agents for adults with type 2 diabetes are unclear. Contribution This systematic review of comparative trials in adults with type 2 diabetes found that premixed insulin analogues and premixed human insulin provided similar glycemic control. Premixed analogues provided tighter glycemic control and caused more hypoglycemia than long-acting insulin analogues and noninsulin antidiabetic agents. Caution Evidence for effects on clinical outcomes was scant and inconclusive. Implication We need large, long-term trials that compare premixed insulin analogues with other agents to see whether improvements in glucose control lead to improved clinical outcomes. The Editors According to the National Health Interview Survey, 28% of patients with type 2 diabetes are using insulin, either alone (16%) or in combination with oral antidiabetic agents (12%) (1). In the management of type 2 diabetes, the role of premixed insulin analogues relative to other insulin regimens and noninsulin antidiabetic agents is unclear. Premixed insulin analogues are derived from rapid-acting insulin analogues and consist of a mixture of a rapid-acting insulin analogue and its intermediate-acting protaminated form. Premixed insulin analogues may be a better alternative than premixed human insulin preparations for patients who wish to have a near-physiologic insulin administration regimen but want to avoid multiple daily insulin injections. In addition, they may allow patients flexible meal times, because these preparations can be administered from 15 minutes before meals to immediately after a meal. Given the increasing prevalence of type 2 diabetes (2), the number of patients who use insulin for glycemic control (1), and the importance of glycemic control in decreasing mortality and morbidity (3), it is imperative to establish the weight of evidence for the safety and effectiveness of these relatively newer insulin preparations compared with traditional insulin preparations. Therefore, the Agency for Healthcare Research and Quality commissioned a systematic review of published studies on the comparative effectiveness and safety of all the premixed insulin analogues that are approved by the U.S. Food and Drug Administration and are available in the United States. Methods Data Sources We searched MEDLINE, EMBASE, the Cochrane Central Register of Controlled Trials, and CINAHL from inception to February 2008. The complete search strategy is available at effectivehealthcare.ahrq.gov/. We also reviewed reference lists of included articles, recent issues of 13 medical journals, the U.S. Food and Drug Administration and European Medicines Agency Web sites for the premixed insulin analogues, unpublished data from premixed insulin analogue manufacturers (Eli Lilly and Company, Indianapolis, Indiana; Sanofi-Aventis, Bridgewater, New Jersey; and Novo Nordisk, Bagsvaerd, Denmark), and Web sites of public registries of clinical trials (ClinicalTrials.gov and ClinicalStudyResults.org). Study Selection We included studies that compared a premixed insulin analogue approved by the U.S. Food and Drug Administration as of February 2008 with any other drug for adults with type 2 diabetes and evaluated clinical outcomes (such as mortality), intermediate outcomes (such as hemoglobin A1c level), or adverse events (such as hypoglycemia). We included randomized, controlled trials (RCTs); controlled clinical trials; and observational studies with control groups, regardless of their duration or sample size. However, we used data from crossover studies only for intermediate outcomes and hypoglycemia. We excluded crossover trials from the quantitative evaluation of outcomes that were either progressive (for example, retinopathy) or irreversible (for example, death). For the evaluation of hemoglobin A1c, we included crossover trials with at least 12 weeks of follow-up before and after the crossover phase. We aimed to use within-individual comparisons from crossover trials if trials had reported data in such detail, but no study did so. Because all crossover studies reported results for each intervention and no trial reported a statistically significant carryover effect, we ignored the crossover design and used reported estimates as if they came from a parallel trial. We excluded nonEnglish-language articles, editorials, comments, letters, and abstracts. Data Extraction and Quality Assessment Two investigators independently reviewed the titles, abstracts, and full articles for inclusion and abstracted data by using standardized forms. We developed a study quality assessment tool based on the Jadad criteria (4), the Newcastle-Ottawa Scale (5), and questions from Agency for Healthcare Research and Qualitys guide for conducting comparative effectiveness reviews (6). We adapted the evidence grading scheme recommended by the GRADE (Grading of Recommendations Assessment, Development and Evaluation) Working Group (7) to classify the strength of the body of evidence on each comparison as high, moderate, low, or insufficient. Data Synthesis and Analysis We conducted meta-analyses for outcomes when data were sufficient (2 trials). For intermediate outcomes (fasting glucose, postprandial glucose, and hemoglobin A1c levels) and the adverse outcome of weight gain, we recorded the mean difference between groups, along with its measure of dispersion. If this was not reported, we calculated the point estimate by using the mean difference from baseline for each group. If the mean difference from baseline was not reported, we calculated this from the baseline and final values for each group. If no measure of dispersion was reported for the between-group difference, we calculated it by using the sum of the variances for the mean difference from baseline for each group. If there were no measures of dispersion for the mean difference from baseline for each group, we calculated the variance by using the standard deviation of the baseline and final values, assuming a correlation between baseline and final values of 0.5. We pooled the results of the plasma and blood glucose levels from different studies because blood glucose measurements accurately reflect plasma glucose levels (8). For hypoglycemia, we used 2 strategies to synthesize data. If a trial reported the incidence of hypoglycemia, we calculated an odds ratio (OR) by using the incidence of hypoglycemia in each study group. If a trial did not report the incidence of hypoglycemia but reported event rates in episodes per patient per 30 days, we calculated the rate ratio by dividing the event rate in the premixed insulin analogue group by the event rate in the comparator group. If a trial reported the number of episodes in each group or reported an event rate in a form other than episodes per patient per 30 days, we converted this information into episodes per patient per 30 days. We pooled the results of individual studies by using a random-effects model. These analyses were conducted by using Comprehensive Meta-Analysis, version 2.2.046 (Biostat, Englewood, New Jersey). For clinical outcomes, we included all studies that reported any information about clinical events (all-cause mortality and cardiovascular mortality and morbidity). All analyses followed the intention-to-treat principle. We combined results from the premixed insulin analogue group of different trials, assuming that the results were similar enough between premixed insulin analogues. In the study with 3 groups and comparing a premixed insulin analogue with 2 different insulin preparations (9), we chose the most relevant comparison to include in the meta-analyses (premixed insulin analogue vs. long-acting insulin analogue). We calculated pooled ORs and 95% CIs by using a MantelHaenszel fixed-effects model (with a 0.1 continuity correction) in Stata Intercooled, version 9.2 (Stata, College Station, Texas) (10, 11). For analysis of clinical outcomes, we used a fixed-effects model because it is less biased with rare event data (12). For sensitivity analyses, we used 3 meta-analytic methods: the Peto method, the MantelHaenszel fixed-effects model (with 0.5 and 0.01 continuity corrections), and Bayesian analysis (13). Heterogeneity among the trials was tested by using a standard chi-square test, with a significance level of 0.10 or less. We also examined inconsistency among studies by using an I 2 statistic (14); a value greater than 50% represented substantial variability. For all outcomes, we conducted sensitivity analyses by omitting 1 study at a time. We assessed publication bias visually by examining the symmetry of funnel plots and statistically by using the Begg (15) and Egger (16) tests. Role of the Funding Source The Agency for Healthcare Research and Quality suggested the initial questions and provided copyright release for this manuscript but did not participate in the literature search, data analysis, or interpretation of the results. Results Study Characteristics The Appendix Figure shows the results of the literature search. We found 45 studies that reported at least 1 of the intermediate clinical outcomes or adverse events (Appendix Table). All studies except 2 (17, 18) were RCTs. In 1 study (17), patients were enrolled consecutively and followed prospectively, and in the other study (18), data were obtained from the medical record database of a large employer. Among the RCTs, 23 were parallel-group (9, 1940) and 20 were crossovers (4160). The median duration of follow-up in these trials was 16 weeks (range, 1 day to 2 years). Appendix Table. Characteristics of the Included Studies Appendix Figure. Study flow diagram. FDA = U.S. Food and Drug Administration. * The total may exceed the number in the corresponding box because articles could be excluded for more than 1 reason at this level. The trials enrolled a total of 14603 patients (median per trial, 93 patient

A meta-analysis and genome-wide association study of platelet count and mean platelet volume in african americans.

Several genetic variants associated with platelet count and mean platelet volume (MPV) were recently reported in people of European ancestry. In this meta-analysis of 7 genome-wide association studies (GWAS) enrolling African Americans, our aim was to identify novel genetic variants associated with platelet count and MPV. For all cohorts, GWAS analysis was performed using additive models after adjusting for age, sex, and population stratification. For both platelet phenotypes, meta-analyses were conducted using inverse-variance weighted fixed-effect models. Platelet aggregation assays in whole blood were performed in the participants of the GeneSTAR cohort. Genetic variants in ten independent regions were associated with platelet count (N = 16,388) with p<5×10−8 of which 5 have not been associated with platelet count in previous GWAS. The novel genetic variants associated with platelet count were in the following regions (the most significant SNP, closest gene, and p-value): 6p22 (rs12526480, LRRC16A, p = 9.1×10−9), 7q11 (rs13236689, CD36, p = 2.8×10−9), 10q21 (rs7896518, JMJD1C, p = 2.3×10−12), 11q13 (rs477895, BAD, p = 4.9×10−8), and 20q13 (rs151361, SLMO2, p = 9.4×10−9). Three of these loci (10q21, 11q13, and 20q13) were replicated in European Americans (N = 14,909) and one (11q13) in Hispanic Americans (N = 3,462). For MPV (N = 4,531), genetic variants in 3 regions were significant at p<5×10−8, two of which were also associated with platelet count. Previously reported regions that were also significant in this study were 6p21, 6q23, 7q22, 12q24, and 19p13 for platelet count and 7q22, 17q11, and 19p13 for MPV. The most significant SNP in 1 region was also associated with ADP-induced maximal platelet aggregation in whole blood (12q24). Thus through a meta-analysis of GWAS enrolling African Americans, we have identified 5 novel regions associated with platelet count of which 3 were replicated in other ethnic groups. In addition, we also found one region associated with platelet aggregation that may play a potential role in atherothrombosis.

Identification of a specific intronic PEAR1 gene variant associated with greater platelet aggregability and protein expression.

Genetic variation is thought to contribute to variability in platelet function; however, the specific variants and mechanisms that contribute to altered platelet function are poorly defined. With the use of a combination of fine mapping and sequencing of the platelet endothelial aggregation receptor 1 (PEAR1) gene we identified a common variant (rs12041331) in intron 1 that accounts for ≤ 15% of total phenotypic variation in platelet function. Association findings were robust in 1241 persons of European ancestry (P = 2.22 × 10⁻⁸) and were replicated down to the variant and nucleotide level in 835 persons of African ancestry (P = 2.31 × 10⁻²⁷) and in an independent sample of 2755 persons of European descent (P = 1.64 × 10⁻⁵). Sequencing confirmed that variation at rs12041331 accounted most strongly (P = 2.07 × 10⁻⁶) for the relation between the PEAR1 gene and platelet function phenotype. A dose-response relation between the number of G alleles at rs12041331 and expression of PEAR1 protein in human platelets was confirmed by Western blotting and ELISA. Similarly, the G allele was associated with greater protein expression in a luciferase reporter assay. These experiments identify the precise genetic variant in PEAR1 associated with altered platelet function and provide a plausible biologic mechanism to explain the association between variation in the PEAR1 gene and platelet function phenotype.

A Meta‐Analysis of the Effect of Thiazolidinediones on Blood Pressure

In epidemiologic studies, insulin resistance is associated with hypertension. Thiazolidinediones (TZDs) are antidiabetic agents that decrease insulin resistance. Multiple clinical trials have evaluated the effect of TZDs on blood pressure (BP) with inconsistent results. The aim of this study was to estimate the effect of TZDs on BP. The authors searched PubMed for clinical trials published in English. A total of 37 clinical trials that reported a change in BP were included in the analysis. Trials with independent‐group design and trials with pre‐post design were evaluated separately. When compared with baseline, TZDs lowered systolic BP by 4.70 mm Hg (95% confidence interval, −6.13 to −3.27) and diastolic BP by 3.79 mm Hg (95% confidence interval, −5.82 to −1.77). When compared with placebo, TZDs lowered systolic BP by 3.47 mm Hg (95% confidence interval, −4.91 to −2.02) and diastolic BP by 1.84 mm Hg (95% confidence interval, −3.43 to −0.25). Thus, TZDs lower both systolic and diastolic BP, albeit the BP‐lowering effect is small and may not be of clinical significance.

The reported validity and reliability of methods for evaluating continuing medical education: A systematic review

Purpose To appraise the reported validity and reliability of evaluation methods used in high-quality trials of continuing medical education (CME). Method The authors conducted a systematic review (1981 to February 2006) by hand-searching key journals and searching electronic databases. Eligible articles studied CME effectiveness using randomized controlled trials or historic/concurrent comparison designs, were conducted in the United States or Canada, were written in English, and involved at least 15 physicians. Sequential double review was conducted for data abstraction, using a traditional approach to validity and reliability. Results Of 136 eligible articles, 47 (34.6%) reported the validity or reliability of at least one evaluation method, for a total of 62 methods; 31 methods were drawn from previous sources. The most common targeted outcome was practice behavior (21 methods). Validity was reported for 31 evaluation methods, including content (16), concurrent criterion (8), predictive criterion (1), and construct (5) validity. Reliability was reported for 44 evaluation methods, including internal consistency (20), interrater (16), intrarater (2), equivalence (4), and test–retest (5) reliability. When reported, statistical tests yielded modest evidence of validity and reliability. Translated to the contemporary classification approach, our data indicate that reporting about internal structure validity exceeded reporting about other categories of validity evidence. Conclusions The evidence for CME effectiveness is limited by weaknesses in the reported validity and reliability of evaluation methods. Educators should devote more attention to the development and reporting of high-quality CME evaluation methods and to emerging guidelines for establishing the validity of CME evaluation methods.

Effect of raloxifene therapy on venous thromboembolism in postmenopausal women. A meta-analysis.

Raloxifene, a selective estrogen receptor modulator, is indicated for the prevention of osteoporosis in postmenopausal women. However, its effect on the risk of deep venous thrombosis (DVT) and pulmonary embolism (PE) is unclear. Therefore, we conducted a meta-analysis to evaluate the effect of raloxifene on these outcomes. To identify randomized controlled trials of raloxifene, a systematic search of PubMed, EMBASE, and Cochrane Collaboration databases was performed from the date of inception of these databases to October 2007. Search was limited to trials that were published in peer-reviewed English-language medical journals. Articles were included in the meta-analysis if they had reported on DVT, PE, or thromboembolic events. Nine trials, including 24,523 postmenopausal women, (median age 59.4 years, range 55 to 67 years; median follow-up 24 months, range 3 to 67 months) met inclusion criteria. Therapy with raloxifene was associated with a 62% increase in odds of either DVT or PE (odds ratio = 1.62; 95% confidence interval = 1.25 to 2.09; p-value < 0.001). Similarly, raloxifene therapy was associated with 54% increase in odds of DVT (odds ratio = 1.54; 95% confidence interval = 1.13 to 2.11; p-value = 0.006) and 91% increase in odds of PE alone (odds ratio = 1.91;95% confidence interval = 1.05 to 3.47; p-value = 0.03). Raloxifene increases the risk of DVT and PE in postmenopausal women.

Platelet Inhibition by Aspirin 81 and 325 mg/day in Men Versus Women Without Clinically Apparent Cardiovascular Disease

Compared with men, women have greater platelet aggregation before and after low-dose aspirin. It is not known whether high-dose aspirin therapy brings residual platelet aggregation in women closer to that in men. Our objective was to compare inhibition of platelet aggregation in women and men after low- and high-dose aspirin. We enrolled healthy subjects (n=106) in a trial of 14 days of aspirin 81 mg/day followed by 14 days of 325 mg/day. Platelet function was measured at baseline and after the 2 aspirin doses. Women had greater baseline platelet activation measurements. After the 2 aspirin doses, men and women had near complete suppression of platelet aggregation to arachidonic acid in whole blood and in platelet-rich plasma (PRP), the direct cyclo-oxygenase-1 pathway affected by aspirin. For indirect pathways, women had significantly greater residual platelet activation to collagen and adenosine diphosphate (ADP) in whole blood after the 2 aspirin doses and in response to collagen and ADP in PRP after aspirin 325 mg/day only. After aspirin 325 mg/day, women continued to have greater residual platelet aggregation compared with men after aspirin 81 mg/day in response to collagen (p=0.016 in whole blood, p=0.037 in PRP), ADP (p<0.001 in whole blood, p=0.012 in PRP), and epinephrine (p=0.03 in PRP). Excretion of urinary thromboxane metabolite (urinary 11-dehydrothromboxane B2) decreased after aspirin to a similar extent in men and women. In conclusion, women continue to have greater residual platelet activity after high-dose aspirin compared with men treated with a lower dose of aspirin.

Effect of Obesity on Platelet Reactivity and Response to Low‐Dose Aspirin

Insufficient platelet function suppression by aspirin is a predictor of cardiovascular events in high-risk patients. The authors assessed the impact of obesity on platelet responsiveness before and after 2 weeks of aspirin 81 mg/d in 2014 people. Obese individuals had greater baseline platelet reactivity. Comparing obese and nonobese individuals after aspirin therapy, results for aggregometry to collagen were 6.7 vs 6.1 ohms, P=.008; aggregometry to adenosine diphosphate were 13.1 vs 11.8 ohms,P<.0001; aggregometry to arachidonic acid (AA) were 4.9% vs 8.3% nonzero aggregation, P=.002; urinary excretion of 11-dehydro-thromboxane B2 (Tx-M) were 4.9% vs 8.3% nonzero aggregation, P=.002; and aspirin resistance were 26.% vs 20.5%, P=.002; respectively. These remained significantly different for AA aggregation and Tx-M excretion after adjustment for covariates. Obese individuals have greater native platelet reactivity and retain greater reactivity after suppression by aspirin.