ライフサイエンスコーパス: censor

1 as 14.0 years; 64% of patients were alive at censor.

2 sets 3-5 years after smoking assessment were censored.

3 , while appropriately treating nondetects as censored.

4 the group of patients who are informatively censored.

5 ufficient sample sizes, or when the data are censored.

6 events with collective action potential are censored.

7 a diagnosis of dementia, died, or were right censored.

8 observed and infectious periods may also be censored.

9 ability weighting to control for informative censoring.

10 used to account for treatment switching and censoring.

11 o adjust adherence estimates for informative censoring.

12 estimated after taking into account interval censoring.

13 on rates between couples and by inconsistent censoring.

14 rculosis diagnosis, death, or administrative censoring.

15 e time of cancer diagnosis or administrative censoring.

16 e they demonstrate two different reasons for censoring.

17 ability weighting to control for informative censoring.

18 DFS: 3-way ITT Pvalue for interaction = .07; censored = .02; IPW = .03; OS ITT Pvalue for interaction

19 = .03; OS ITT Pvalue for interaction = .007; censored = .04; IPW = .03).

20 Data for all patients were censored 1 year post-study or death, whichever came firs

21 Whereas 1975 patients received a KT and were censored, 1876 were on the waiting list at any time.

22 Performance was calculated after censoring 365 days after prior screen, with modeling of

23 At time of data censoring, 6 of 11 patients had not progressed.

24 Death-censored actuarial 15-year graft survival rate was 56%.

25 t a median follow-up of 31 months, the death-censored actuarial graft survival of dDSA recipients was

26 rt was 31 mo (interquartile range 22-46 mo), censored after 26 October 2013.

27 od to calculate the cumulative risk of death-censored allograft failure may overestimate the risk of

28 Secondary outcomes included death and death-censored allograft failure.

29 core has recently predicted 50%10-year death-censored allograft loss in patients with donor-specific

30 ifference between patient survival and death-censored allograft survival (graft survival).

31 The 1-, 3-, 5- and 7-year death-censored allograft survival rates were 98%, 91%, 86%, an

32 ith a cAMR score less than 13, 10-year death-censored allograft survival was 96% to 100% regardless o

33 Death-censored allograft survival was similar in all groups ex

34 Death-censored AMR-free and allograft survivals were significa

35 y outcomes were compared in paired, pairwise-censored analyses.

36 Death-censored analysis demonstrated no increased relative ris

37 Using a prespecified lag-censoring analysis (a measure of actual drug exposure),

38 h a prior history of any MACE before MI were censored and adjusted for follow-up times.

39 Kaplan-Meier estimates for both, patient-censored and death-censored graft survivals were both 97

40 Analyses adjusting for crossover (censored and inverse probability weighted [IPW]) were al

41 , and prior acute rejection) predicted death-censored and overall graft survival (c statistics =0.84

42 proportion of patients who are informatively censored and secondarily on the hazard ratio between the

43 Overall, death-censored and technically successful pancreas graft survi

44 ldwide network of computers which texts were censored and which were not.

45 By dealing with censoring and competing risk of death, we developed a sc

46 of an illness-death model handling interval censoring and competing risk of death.

47 or age and follow-up time, and accounted for censoring and competing risk of death.

48 estimated from epidemiologic data subject to censoring and competing risks with adjustment for multip

49 probabilistic approach that accounts for the censoring and evaluate it for two typical datasets conta

50 To account for right-censoring and interval censoring, we estimated the ROC c

51 entration responses reveals a combination of censoring and mapping the fluorescence responses to conc

52 unlike concentration values, doesn't require censoring and we show with respect to differential analy

53 for (a) mortality, (b) allograft loss (death censored), and (c) combined death or transplant failure.

54 CI 81-95; 114 total patients, 12 events, 36 censored), and 27 (22%) of 121 patients died by the end

55 as implications, thus we recommend using the censoring approach for event rate estimation among AF pa

56 the future' approach, and 0.93 when using a 'censoring approach'.

57 for noise models and kernels accounting for censoring are available at http://icb.helmholtz-muenchen

58 spleen where autoreactive specificities are censored as B cells gain immune competence, but the intr

59 come was the time until lung transplantation censored at 1 year.

60 lculated overall hazard ratios for mortality censored at 14 days using Cox proportional hazards model

61 the primary endpoint of time-to-death right-censored at 2 years.

62 When censored at 3 months, CompEx resulted in 2.8 times more

63 tcomes were obtained via Medicare claims and censored at 3 years.

64 tudy endpoints, including hospital mortality censored at 30 days and clinical cure.

65 iffuse lung diseases, and chose likelihoods (censored at 5% and summing to 100% in each case) for eac

66 The RSE, right censored at 6 months to include all 3 studies, was based

67 ollowed up patients until hospital discharge censored at 60 days, estimated incidence from prevalence

68 ntion-to-treat population (n=675), with data censored at crossover.

69 r death after graft failure; observation was censored at death with graft function.

70 Time-to-event analyses were censored at first date of new cancer event, last contact

71 Patients were censored at postdischarge emergency department encounter

72 bimatoprost-treated pooled fellow eyes (data censored at rescue/retreatment).

73 nts who received SCT in first remission were censored at SCT time, 2-year RFS was 53.3% (95% CI, 39%

74 5 patients received a successful KT and were censored at that point, whereas 1876 were on the waiting

75 e variable; asymptomatic carriers were right censored at the age at last contact or age at death.

76 cipants were followed up for at least 2 y or censored at the last follow-up.

77 ts All patients in the N3I3 arm (n = 6) were censored at the time of analysis as a result of dose-lim

78 When participants were censored at the time of cointerventions (parathyroidecto

79 nts undergoing aortic valve replacement were censored at the time of surgery (n=92).

80 tes; children living on March 31, 2013, were censored at their last clinical encounter.

81 who continued, alive and relapse free, were censored at their last known follow-up.

82 e group had been lost to follow-up, and were censored at their last visit for the primary analysis.

83 Survival at 12 months after listing (censored at Tx) was worse in patients on ECMO at listing

84 n until death from any cause, administrative censoring at 10 years after therapy initiation or the en

85 into care until death, loss to follow-up, or censoring at 5 years or on December 31, 2013.

86 iate Cox regression analyses were performed, censoring at cardiac transplantation, to assess the impa

87 ene expression more effectively than classic censoring-based approaches and leads to power gains in d

88 2 y or until occurrence of a CRBSI or right-censoring because of CVC removal.

89 Data were censored both at the time of transplantation (listed onl

90 996 and 2012 and followed up through 2013 or censored by death or emigration.

91 ded as serodiscordant only once before being censored by loss to follow-up, couple dissolution, or st

92 Our proposed methodology handles the "censoring by death" phenomenon through principal stratif

93 tent variable model framework to account for censoring by introducing an appropriate noise model and

94 Because this censoring can occur for high proportions of the data, it

95 If informative censoring cannot be avoided, then all patients should be

96 incidence of the composite outcome or death-censored cardiovascular events over time (P = 0.41 and 0

97 After censoring cases in which patients died during the follow

98 abetes mellitus or end of follow-up owing to censoring caused by the transition into a Medicaid manag

99 a 3D genome organization to both promote and censor communication along and between chromosomes.

100 on effects of hypothetical interventions and censor competing events.

101 Death-censored cumulative events were analyzed using Kaplan-Me

102 s was tested using the concordance index for censored data and decision curve analysis.

103 By utilizing right-censored data in its training process, the method demons

104 ssion models for left-truncated and interval-censored data to simultaneously estimate the association

105 Quantitative challenges presented by censored data were addressed with nonparametric distribu

106 50 data sets in which the true values of the censored data were known, and therefore "true" probabili

107 use of linear mixed effect models including censored data, thereby considering data below detection

108 a parametric survival analysis for interval-censored data.

109 date it is not clear how to perform a PCA of censored data.

110 Lognormal models and censored-data methods produced estimates of chemical ass

111 andomisation and death from any cause or the censor date) in the intention-to-treat population.

112 The censoring date was May 31, 2014.

113 The interim analysis of OS (64% censored) demonstrated a 13% reduction in risk of death

114 nurse to the occurrence of an outcome, or to censoring due to completion of service or the end of ava

115 For higher degrees of censoring (each group >40% censoring), no technique prov

116 have been used to determine if two groups of censored environmental data arise from the same distribu

117 re followed up for at least 1 year, or had a censoring event.

118 The midpoint approximation for interval-censored exposures led to overestimation of the mean inc

119 uding adjusting for plasma viral load assay, censored follow-up at 3 years, or used variations of the

120 We had prespecified an analysis censoring follow-up at oral poliovirus vaccine campaigns

121 Participants were censored for a new diagnosis of a uveitis-associated sys

122 = .779; Q = 13.03, I(2) = 38.6%), graft loss censored for death (OR = 0.73; 95% CI, .17-3.21; P = .67

123 that significantly influenced graft failure censored for death was peripheral vascular disease.

124 In multivariable analysis, graft survival censored for death was significantly influenced by recip

125 Patients were censored for death with LVAD at the time of transplant o

126 dent influence on the risk of graft failure, censored for death.

127 ival (PFS) analysis arises when patients are censored for initiation of an effective anticancer treat

128 If censored for primary nonfunction, estimated glomerular f

129 rred (822 cardiovascular), with 655 patients censored for renal transplantation and 1183 for loss to

130 r age, sex, and cardiovascular risk factors; censored for stroke; and stratified by median age were u

131 Censoring for concentration data caused problems for ana

132 , 1.15 to 1.64; P<0.001) that persisted when censoring for death (HR, 1.43; 95% CI, 1.12 to 1.84; P=0

133 4) and a 34% reduction (0.66; .44-1.00) when censoring for oral poliovirus vaccine campaigns.

134 The pretransplant population was censored from further survival analysis on receipt of a

135 ths; GFNC represented 48.1% (26/54) of death-censored GFs beyond 24 months.

136 transplants performed at 101 centers, death-censored graft and patient survival rates were similar t

137 Death-censored graft and patient survival were not significant

138 HR]: 1.01, 95% CI: 0.98-1.04, P = .4), death-censored graft failure ( [aHR]: 1.02, 95% CI, 0.98-1.06,

139 interval [CI], 1.36-2.32), and 3 month death-censored graft failure (adjusted hazard ratio, 2.0; 95%

140 ersus late events on risk of long-term death-censored graft failure (DCGF).

141 erval [95% CI], 1.30-2.35; P < 0.001), death censored graft failure (hazard ratio [HR], 2.06; 95% CI,

142 confidence interval, 1.87 to 2.60) and death-censored graft failure (hazard ratio, 5.14; 95% confiden

143 e (HR [95% CI], 1.97 [1.09-3.56]), and death-censored graft failure (HR [95% CI], 2.43 [1.07-5.51]).

144 -1.2; P < 0.001) and a similar risk of death-censored graft failure (HR,1.0, 95% CI, 1.0-1.1; P = 0.1

145 Primary outcome measures were death-censored graft failure and all-cause mortality.

146 inition, 14 (3%) died, and 23 (4%) had death-censored graft failure by 12 months.

147 However, the relative risk of death-censored graft failure of a 2-2-2 mismatched living-dono

148 late TG for the composite endpoint of death-censored graft failure or doubling of serum creatinine.

149 models to determine risks of death or death-censored graft failure related to percentage change in e

150 79.8+/-5.9 at 60 months, P=0.01), and death-censored graft failure was significantly higher in group

151 ted with risks of subsequent death and death-censored graft failure, which mirrors findings in CKD.

152 association between recipient sex and death-censored graft failure.

153 6) for death, and 0.62 (0.49-0.78) for death-censored graft failure.

154 FNC, representing 67.6% (25/37) of its death-censored graft failures observed beyond 24 months after

155 ; 95% CI, 1.38-1.83, respectively) and death-censored graft loss (aHR, 1.41; 95% CI, 1.24-1.60 and aH

156 with a substantially increased risk of death-censored graft loss (aHR, 5.24; 95% CI, 1.97-13.98, P =

157 nter variation exists in mortality and death-censored graft loss (DCGL) after transplantation.

158 ations between DGF status, overall and death-censored graft loss (DCGL) were examined using adjusted

159 development and contribute to further death-censored graft loss (DCGL).

160 terval, 1.13-1.88; P < 0.001), but not death-censored graft loss (hazard ratio, 1.25; 95% confidence

161 e (HR, 1.30; 95% CI, 1.07 to 1.58) and death-censored graft loss (HR, 1.41; 95% CI, 1.12 to 1.78).

162 fidence interval [95% CI], 1.07-1.33), death-censored graft loss (HR, 1.67; 95% CI, 1.45-1.92), and l

163 er due to recipient death (n = 540) or death-censored graft loss (n = 353).When the observational tim

164 We estimated the risk of death-censored graft loss and mortality after developing demen

165 d hazard ratio (aHR) of patient death, death-censored graft loss and posttransplant malignancy associ

166 that researchers should focus on using death-censored graft loss as the primary outcome of interest t

167 had experienced overall graft loss and death-censored graft loss at 3 years compared with those witho

168 2.31-7.58, P < .001) by 12 months and death-censored graft loss by 5 years (HR 3.12, 95% CI 1.53-6.3

169 nced a higher incidence of overall and death-censored graft loss compared with those without DGF.

170 dding these variables to the model for death-censored graft loss increased predictability (c statisti

171 Uncensored and death-censored graft loss occurred in 263 and 46 recipients, r

172 In fully adjusted models, only death-censored graft loss remained significant (HR, 1.38; 95%

173 Depression increased death-censored graft loss risk (RR, 1.65; 95% CI, 1.21-2.26, 3

174 Death-censored graft loss was higher with late versus early AM

175 interval [CI], 0.93-1.20]; P = 0.40), death-censored graft loss was lower (HR, 0.63; 95% CI, 0.47-0.

176 Data on death-censored graft loss were obtained from the Norwegian Ren

177 Sixty-six patients (17%) had death-censored graft loss, and 116 (30%) patients died.

178 Primary endpoint was death-censored graft loss, and secondary endpoint was all-caus

179 eated mismatch increased the hazard of death-censored graft loss, particularly in patients with detec

180 of repeated mismatches on all-cause or death-censored graft loss.

181 0.98, P = 0.006) were associated with death-censored graft loss.

182 atrophy (IFTA) on 24-month biopsy and death-censored graft loss.

183 with overall renal graft loss, but not death-censored graft loss.

184 not associated with either overall or death-censored graft loss.

185 n was associated with both overall and death-censored graft loss.

186 equent (P=0.017) and earlier (P=0.046) death-censored graft loss.In addition, daily medication adhere

187 The proportion of death-censored graft losses caused by CRS was 4.6% (7/152 pati

188 tological criteria and associated with death-censored graft outcome.

189 1 year, 89%; P < 0.01), but decreased death-censored graft survival (5 years: preKT, 93%; dialysis <

190 The 8-year death-censored graft survival (DCGS) rate was 56.8% of TCMRV v

191 Up to 8-year posttransplantation, death-censored graft survival (DCGS) rates of control, borderl

192 timated glomerular filtration rate and death-censored graft survival (DCGS).

193 Death-censored graft survival at 3 years was comparable for th

194 Death-censored graft survival at last follow-up was 100% in th

195 Death-censored graft survival in MN did not differ from that o

196 Death-censored graft survival in the R+/D+ population was bett

197 In the death-censored graft survival model, there was no statistical

198 f two SNPs on chromosomes 14 and 18 on death-censored graft survival or all-cause mortality was not c

199 adsorption was associated with similar death-censored graft survival to plasmapheresis.

200 Patient survival was 96.3% (n=52), and death-censored graft survival was 100% at a median follow-up o

201 Death-censored graft survival was 84%.

202 analyzed at 2 and 5 years, the 10-year death-censored graft survival was lower for patients with C1q-

203 tion episodes (P < 0.001), but reduced death-censored graft survival was not observed after a median

204 In multivariate analysis, death-censored graft survival was significantly associated wit

205 ion group, compared with no rejection, death-censored graft survival was significantly worse in 23 pa

206 Death-censored graft survival was significantly worse in those

207 A were analyzed at 2 years, the 5-year death-censored graft survival was similar between patients wit

208 Death-censored graft survival was unaffected.

209 Three-year death-censored graft survival was virtually identical for ABO-

210 t histopathology, rejection rates, and death-censored graft survival were not significantly different

211 aitlist, which contributed to improved death censored graft survival when compared with R+/D- patient

212 Death-censored graft survival, patient survival, and rejection

213 ts (vs primary) had decreased mid-term death censored graft survival.

214 he first posttransplantation week, and death-censored graft survival.

215 rt- and long-term patient survival and death-censored graft survival.

216 The overall graft survivals and death censored graft survivals among groups were not statistic

217 timates for both, patient-censored and death-censored graft survivals were both 97.2% at 5 years.

218 nt survival (PS), graft survival (GS), death-censored GS (DCGS), and acute rejection-free survival (A

219 Outcome data were censored if no CCHS encounters occurred for 2 consecutiv

220 Informative censoring in a progression-free survival (PFS) analysis

221 arametric models accounting for the interval censoring in some exposures were fitted to the data.

222 of a novel statistical method accounting for censoring in the follow-up period to a nationwide twin s

223 cordance, or C statistic, which accounts for censoring in time-to-event models (a-c).

224 alyze age of onset and death in affected and censored individuals.

225 Taking the censoring into account results in a 2D representation of

226 al stratification and inverse-probability-of-censoring (IPC) weights.

227 Death-censored kidney graft survival was nevertheless comparab

228 MR score in a separate cohort predicts death-censored long-term allograft failure in DSA+ patients re

229 s from a commonly occurring form of interval-censored longitudinal parasitological data-specifically,

230 For low degrees of censoring (<25% in each group), the Generalized Wilcoxon

231 dies of incident exposures may involve right censoring (missingness on the right) and therefore may n

232 For moderate degrees of censoring, MLE worked best, but only if the distribution

233 We used multivariable interval-censored models to evaluate associations of OCPs (quarti

234 Multivariable interval-censored models were used to evaluate associations of li

235 higher degrees of censoring (each group >40% censoring), no technique provided reliable estimates of

236 A censored normal regression model provided the best fit m

237 hat result from treating competing events as censored observations and how they relate to measures of

238 e explore the implications of this model for censored observations and the effect on genomic predicto

239 Censoring occurred at uterine cancer diagnosis, hysterec

240 the distribution is complicated by interval censoring of exposures.

241 in limitations of the study include interval censoring of incident dementia cases, potential selectiv

242 nal networks as they often require threshold censoring of information and do not allow for inferentia

243 95% confidence interval, 0.31 to 0.93); with censoring of time after kidney transplantation, the rela

244 pretations are available for H3N2, but right-censoring of titers makes these interpretations difficul

245 Follow-up data were censored on December 13, 2011.

246 atients were followed up until death or were censored on December 31, 2013.

247 new non-recrudescent malaria infection] were censored on the last day of follow-up), and per-protocol

248 eled the distribution of ICC as a mixture of censored or truncated normal and normal distributions us

249 lative to the other techniques regardless of censoring or group size.

250 Variable selection for censored outcome data as well as control of false discov

251 Death-censored outcomes after transplant differed significantl

252 hical Weibull regression model with interval-censored outcomes.

253 or any mental disorder; 489,006 persons were censored owing to death; and 69,987 persons were censore

254 ored owing to death; and 69,987 persons were censored owing to emigration.

255 performed with Cox regression with survival censored past 90 days.

256 Median follow-up in censored patients was 8.3 years.

257 observed after adjustment for age and after censoring patients who received allogeneic stem cell tra

258 ity" was strongly dependent on the degree of censoring present in the groups.

259 3 patients died, and 844 patients are alive (censored rate, 85.0%).

260 Censored regression analysis on all affected and unaffec

261 Multivariate censored regression was used to assess the association b

262 times are not observed, detection times are censored, removal times are known, and the disease is sp

263 999 and 2011, we estimated overall and death-censored renal graft loss hazard ratios in patients diag

264 f an increased risk of overall but not death-censored renal graft loss in renal transplant recipients

265 ollow-up of 19 months, there was 100% (death-censored) renal allograft survival with estimated glomer

266 with propensity score matching and pairwise censoring, respectively.

267 d the variance of the estimator in the right-censoring setting.

268 VEA was associated with ischemic stroke when censoring subjects at time of AF (hazard ratio [HR]: 1.9

269 Using interval-censored survival and binomial regression approaches a m

270 oposed procedure on two cancer datasets with censored survival outcomes and thousands of molecular fe

271 Kaplan-Meier death-censored survival plots and Cox regression were used to

272 distributions were modeled by using interval-censoring survival analysis with 3 parametric approaches

273 cting with Chinese firms to install the same censoring technologies as existing sites, and--with thei

274 When censoring the patients at the time of stent thrombosis,

275 patients who underwent transplantation were censored, the benefit of midostaurin was consistent acro

276 Interval-censored time-updated proportional hazards regression wa

277 The median survival time, censored to liver transplantation, was 17.7 months for t

278 Animal data were censored to simulate prospective monitoring at any momen

279 ked from wait list entry date until death or censoring to determine influence of PH.

280 failure time regression models with interval censoring to estimate time ratios and hazard ratios and

281 behaviors and socioeconomic status, and left-censoring to explore reverse causality had very little i

282 ing (IPW) of bacteremia or sepsis and IPW of censoring, to estimate the marginal causal effects of ba

283 ") contaminant concentrations from data with censored values (e.g., less than the detection limit).

284 nt dementia and its subtypes were studied as censored variables using Cox models with age as time sca

285 ncident mild cognitive impairment (N=365) or censoring variables (N=179) for a median of 5 years.

286 rson-years (n = 23), and prevalence of VF at censoring was 17.8%.

287 Mean follow-up until death or censoring was 4.2 years.

288 Bias due to dependent censoring was investigated via inverse probability weigh

289 Cox regression with left and right censoring was used to estimate cumulative incidence of d

290 Cox regression with left and right censoring was used to estimate the frequencies and relat

291 , customized preprocessing, including volume censoring, was used to minimize motion-induced rs-fcMRI

292 To account for right-censoring and interval censoring, we estimated the ROC curves by means of a wei

293 For rare failure events subject to censoring, we have proposed efficient augmented case-onl

294 Cox models that used inverse probability of censoring weighted modeling.

295 by competing risk and inverse probability of censoring weighting analyses accounting for transplantat

296 We performed inverse probability of censoring weighting and competing risk analyses.

297 We calculated inverse probability of censoring weights to adjust adherence estimates for info

298 matched, and both left-truncation and right-censoring were accounted in order to compare higher IPSS

299 r being HIV-infected on cART, with follow-up censored when cART regimen was modified, was associated

300 Liability threshold models adjusting for censoring with inverse probability weighting were used t