Outcomes Study Short Form-36 (SF-36) (a validated generic health status measure),19 at the third and/or ﬁfth years.20 Statistical analysis Treatment history All analysis was conducted using Stata version 9.2 (StataCorp, College Station, Texas, USA). The time from symptom onset to the date of the ﬁrst treatment was categorised into three groups (<6 months, 6–12 months and >12 months). The duration of the ﬁrst DMARD/S treatment and total duration of treatment over the 10 years of follow-up, taking into account combination therapies, were divided into tertiles (ﬁrst DMARD/S treatment ≤6 months, >6 to ≤36 months and >36 months; total duration ≤48 months, >48 to ≤96 months, >96 months). Outcome by 10 years and treatment The change in HAQ score between baseline and the 10-year follow-up was included in a linear regression model, adjusted by baseline HAQ, in order to estimate the effect on long-term functional disability of treatment, as well as the time to, and time receiving, the ﬁrst treatment, and total time receiving treatment. Patients who had never received DMARD/S were used as the referent group. To explore the effect of the interaction between time to, and time receiving, treatment the duration of the ﬁrst and any DMARD/S treatment, including combination therapies, were stratiﬁed into the number of months of treatment in the ﬁrst 6 months since symptom onset, and the number of months of treatment after the ﬁrst 6 months since symptom onset. Marginal structural models We have previously adjusted for ‘confounding by indica- tion’ using propensity scores.9 10 However, while this method adjusted for initiation of ﬁrst DMARD/S, no adjustment was made for subsequent treatment decisions. On this occasion, we used marginal structural models to adjust for time-dependent confounders, such as the HAQ.8 21 At each follow-up, a logistic regression model was used to estimate the probability that the patient was receiving the observed treatment. Baseline, previ- ous and assessment-speciﬁc variables were used in the model to predict the treatment decisions as outlined in table 1. This method aims to include as many potential predictors as possible, even if their contribution is small. The reciprocal of the prob- ability was used if the patient was not receiving treatment at that assessment. The product of all the probabilities for each patient gave the overall probability of the treatment decisions that occurred over the length of follow-up. The inverse of this probability (the inverse-probability-of-treatment weight) was then used to adjust each patient’s change in HAQ for the vary- ing treatments and time-dependent confounders. However, without a numerator this inverse weight is likely to be highly variable and so the inverse-probability-of-treatment weight is stabilised.21 The numerator was the product over the length of the follow-up of the proportions of patients receiving treatment at each assessment. To allow for loss to follow-up, a stabilised inverse-probability- of-censoring weight was calculated, based on the same methods outlined above. Without adjustment for loss to follow-up, any estimate of the change in HAQ in the tenth assessment com- pleters would be biased. At each assessment, the probability of the subject still being in NOAR by the next assessment was esti- mated using a logistic regression model with the same variables as in the treatment decision model. The inverse product of these probabilities was also stabilised and this censoring weight was multiplied by the stabilised inverse-probability-of-treatment Ann Rheum Dis 2010;69:689–695. doi:10.1136/ard.2009.108639 Extended report weight to provide a weight that produced a pseudo population by the tenth assessment, in which both treatment and loss to follow-up are independent of any of the variables considered to be potential time-varying confounders, enabling an unbiased estimate of the effect of treatment. The logistic regression models can only include patients with complete data. In total, 162 840 data items were included in the models used to produce the weights. Of these, 6785 (4.2%) were missing. To impute missing data we used switching regression, an iterative multivariable regression technique which retains an element of random variation in the estimates.22 Within Stata these methods are incorporated within the ice and uvis programs. For each variable the distributions were compared with and without the imputed data to conﬁrm that the imputed data did not alter the distribution of any of the variables. The imputed data were only used in the prediction of treatment decisions and loss to follow-up, and were not included in the change in HAQ models as the marginal structural model accounts for loss to follow-up as well as time-dependent confounders. RESULTS Cohort characteristics and treatment over 10 years of follow-up The median age at symptom onset of the 1084 subjects was 53 years (IQR 41–66) with two-thirds being women (table 2). At the baseline assessment, 493 (45.5%) subjects satisﬁed the 1987 ACR criteria for RA. After 10 years, 664 (61.3%) subjects continued to be followed up. There was an increasing use of MTX, a decreasing use of sulfa- salazine (SSZ) and a stable use of steroids with time (table 3). At baseline, 7% of subjects receiving a DMARD/S were receiv- ing MTX and 60% SSZ. At the tenth year assessment, 54% were receiving MTX and 37% SSZ. This reﬂects the changing management of RA during the 1990s. When restricted to those subjects who satisﬁed the ACR criteria for RA by 10 years, the proportion treated was higher (eg, 64% ever treated by 10 years) but the pattern of the types of treatment was the same (data not shown). HAQ score at baseline, 10 years and difference over 10 years by time to, and time receiving, treatment Of the 664 subjects with 10 years of follow-up, 642 subjects had completed a HAQ both at baseline and the 10-year assess- ment; 54% of these had been treated by 10 years, of whom 35% were treated within 6 months of symptom onset (table 4). As expected patients who were judged not to require DMARD/S treatment by the tenth assessment had milder disease with lower baseline HAQ scores and smaller mean change in HAQ over the follow-up period (0.13, 95% CI 0.05 to 0.21) than those who were treated. The patients with the highest baseline HAQ were most likely to receive early treatment, but had no mean change in HAQ over the 10 years (0.00, 95% CI –0.17 to 0.17). Those who continued to receive their ﬁrst DMARD the longest, experienced the least deterioration in HAQ over the follow-up period. Patients who continued to receive their ﬁrst DMARD/S for 3 years or more experienced a deterioration in HAQ of only half (0.16, 95% CI 0.002 to 0.32) that seen in patients who stopped their ﬁrst DMARD within 6 months of starting it (0.35, 95% CI 0.15 to 0.55). After an initial improvement in the ﬁrst year of follow-up, all groups showed a steady decline in functional disability over the 10 years (ﬁgure 1). Patients who were receiving DMARD/S treatment at each follow-up reported poorer function than those 691